ARGUMENTATIVE ESSAY (Scholarly work required) I DISPUTE bad work!!
DO NOT BID IF YOU CANT PROVIDE QUALITY WORK AND FOLLOW DIRECTIONS. I WILL DISPUTE FOR REFUND IF BAD WORK IS PROVIDED!!
– Topic- Why Labor Unions are Important in Today’s Workforce (United States)
– Essay can only speak on United States
– 4-5 pages in length
– APA 7th Ed.
– Must use provided references only and utilize minimum (8) citations from them
– Must use appropriate Headings in APA 7th Ed format
– Must have Opening Thesis statement clearly defined
– Must have supporting statements
– Must have counter argument
– Must have Clear conclusion supporting topic of why labors unions are important
** See attachments for additional information and reference/research material
References
Supporting Points:
a. Collective Voice
Lott, B. (2014). Social Class Myopia: The Case of Psychology and Labor Unions. Analyses of Social Issues & Public Policy, 14(1), 261–280.
https://doiorg.nuls.idm.oclc.org/10.1111/asap.12029
b. Equality
Cho, D., & Cho, J. (2011). How do labor unions influence the gender earnings gap?: A comparative study of the US and Korea. Feminist Economics, 17(3), 133–157.
https://doi.org/10.1080/13545701.2011.582472
Flavin, P. (2018). Labor Union Strength and the Equality of Political Representation. British Journal of Political Science, 48(4), 1075–1091.
https://doi.org/10.1017/S0007123416000302
Macdonald, D. (2019). Labor Unions and Support for Redistribution in an Era of Inequality. Social Science Quarterly (Wiley-Blackwell), 100(4), 1197–1214.
https://doi-org.nuls.idm.oclc.org/10.1111/ssqu.12627
c. Sense of belonging
Dawkins, C. (2016). A Test of Labor Union Social Responsibility: Effects on Union Member Attachment. Business & Society, 55(2), 214–245. https://doi.org/10.1177/0007650312464925
Counter Arguments:
Clawson, D., & Clawson, M. A. (1999). What has happened to the US labor movement? union decline and renewal. Annual Review of Sociology, 25, 95-119. Retrieved from
https://nuls.idm.oclc.org/login?url=https://search-proquest-com.nuls.idm.oclc.org/docview/199590715?accountid=25320
Raymo, J., Warren, J., Sweeney, M., Hauser, R., & Ho, J. (2011). Precarious employment, bad jobs, labor unions, and early retirement. The Journals of Gerontology. Series B, Psychological Sciences and Social Sciences, 66(2), 249–259. https://doi.org/10.1093/geronb/gbq106
1
Outline for Argument Essay
Title: The Need for Labor Unions in Today’s Workforce.
A 4 to 5-page argument paper that requires the use of scholarly, trade journal and textbook sources.
I. Introduction of the Essay
A. Hook: State your position and give an interesting angle to support your view.
B. End this section with a thesis sentence – see common student mistakes for help.
II. Body of the Paper
A. General Background on Topic (Choose at least three topics – and ensure you are not only using one sided argument. What does the other side say? Make sure you counter this argument):
B. General Background on Topic B:
C. Identify similarities between Topic A and Topic B
D. Identify differences between Topic A and Topic B
E. Analysis of the topict: There are (few, many, several) similarities and (few, many, several) differences between _____________________________________ and _______________discuss your final outcome by giving evidence to support your point of view_______.
III. Conclusion
A. Restatement of main points.
B. Insights into the issues.
C. Strong Ending: Which side are you on? By the end, there should be overwhelming evidence to support your selection. This includes countering the other sides arguments too.
Business & Society
2016, Vol. 55(2) 214 –245
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1California State Polytechnic University, Pomona, CA, USA
Corresponding Author:
Cedric E. Dawkins, Rowe School of Business, Dalhousie University, 6100 University Avenue,
Halifax, NS B2Y0C1 Canada.
Email: dawkins@dal.ca
The article was accepted during the editorship of Duane Windsor.
A Test of Labor Union
Social Responsibility:
Effects on Union Member
Attachment
Cedric E. Dawkins1
Abstract
Social responsibility is addressed to corporations, but can also be applied
to other powerful organizations. This study tests the impact of labor union
social responsibility on key measures of labor union attachment. After devel-
oping a scale of labor union social responsibility, craft union apprentice work-
ers were surveyed and their responses analyzed with structural equation
modeling. Labor union social responsibility was directly and positively related
to union commitment and job satisfaction. Union commitment and job sat-
isfaction fully mediated the negative relationship between labor union social
responsibility and propensity to withdraw from the union, and the positive
relationship between labor union social responsibility and union participa-
tion. The results suggest that labor union social responsibility can enhance
union attachment and inform union strategy.
Keywords
labor unions, social responsibility, union member attachment
In the global economy the notion of social responsibility, defined as actions
that further some social good beyond self-interest and what is required by law
Article
http://crossmark.crossref.org/dialog/?doi=10.1177%2F0007650312464925&domain=pdf&date_stamp=2012-11-01
Dawkins 215
(McWilliams & Siegel, 2001), has gained currency among stakeholders (Doh
& Guay, 2006; Matten & Crane, 2005) and academics (Margolis & Walsh,
2003). Consumers expect social responsibility (Mohr & Webb, 2005; Mohr,
Webb, & Harris, 2001), and the public rewards socially responsible compa-
nies with enhanced reputation (Fombrun, Gardberg, & Barnett, 2000) and
employee recruiting preferences (Turban & Greening, 1997). Social respon-
sibility expectations have been primarily directed toward businesses, but
labor unions are also very powerful organizations with broad societal influ-
ence. As providers of essential goods and services, unions have profound and
widespread impact on consumers (e.g., education, public transportation,
health care, first responder services, and trucking), and play a major role in
the economic well being of their members, communities, and countries.
Hence, there is little reason for unions to escape the imperative to operate in
a socially responsible manner.
To date, labor unions have been somewhat conflicted about corporate
social responsibility (CSR). Despite historically striving for what many
would argue are key tenets of corporate social responsibility—an equitable
wage, humane working conditions, due process for workers, and rights for
marginalized communities1—some union leaders fear that social responsibil-
ity will undermine their preferred structure of contracts and regulation
(Justice, 2003; Preuss, 2008). Others contend that a broad, socially conscious
labor movement that is genuinely concerned about social justice is better
positioned for the future (FNV Mondiaal, 2004; Wheeler, 2001), and that a
more favorable view of unions by members and potential members is a req-
uisite for union revitalization (Freeman & Rogers, 1999). Amid this reti-
cence, U.S. labor unions are beset with the public perception that they
selfishly pursue the interests of their members (Panagopoulos & Francia,
2008), and skepticism about their purpose and power (Pew Research Center,
2010). This article espouses the view that social responsibility is itself a valu-
able end. If, however, there are connections between labor union social
responsibility (LUSR) and key measures of union attachment, then identify-
ing the nature of those connections can inform discussion of union revitaliza-
tion, strategy, and tactics.
There is a paucity of research that explores the nexus of labor unions and
social responsibility, but LUSR has potentially important implications for
union members and the societies to which they belong. This study focuses on
union members’ perceptions of LUSR and the extent to which those percep-
tions impact union attachment. The first section describes why labor union
social responsibility is warranted, its benefactors, and how labor unions can
discharge social responsibility. The second section presents hypotheses
216 Business & Society 55(2)
regarding the extent to which perception of LUSR positively impacts key
measures of union attachment such as union commitment, job satisfaction,
propensity to participate, and propensity to withdraw. The third section pro-
vides an explanation of the methodology of the study, participants and proce-
dure; measures, and the analytic strategy for structural equation modeling
and hypothesis testing. After presenting the results of hypothesis tests, the
article concludes with discussion of the implications of LUSR for business
ethics and labor relations research and practice.
Labor Union Social
Responsibility
Dawkins (2010) describes socially responsible labor unions as those that
pursue economic equity, workplace democracy, and social justice in ways
that are consistent with general expectations of organizational conduct, con-
sider the interests of their stakeholders, and benefit society. A discussion of
LUSR can begin with three basic questions: Why is labor union social
responsibility warranted, to whom are labor unions responsible, and how can
labor unions discharge their responsibility?
Why labor union social responsibility is warranted. Concerning the first ques-
tion, labor union social responsibility is warranted because the actions of labor
unions can, and often do, affect or put at risk the interests of others. Social
responsibility implies that powerful organizations occupy a role that extends
beyond laws and regulations to encompass a wide range of societal norms,
values, and expectations (Carroll, 1991). Support for this supposition is derived
from institutional theory, deontological ethics, and social contract theory.
According to institutional theory, organizations whose actions are consis-
tent with the normative values of society are deemed legitimate and receive
preference in transacting business (DiMaggio & Powell, 1983; Suchman,
1995). That preference is, however, constrained by responsibility because it
might otherwise produce negative externalities (Valasquez, 1996). The iron
law of responsibility states that businesses are social institutions that must use
their power responsibly, or risk losing that power altogether (Davis &
Blomstrom, 1971). Through collective bargaining labor unions play a major
role in domestic economic health, function as the sole representative of their
members’ workplace interests,2 and provide essential goods and services for
consumers. It follows that if labor unions operate in a manner that is incon-
sistent with the goals and values of society they are also likely to lose their
societal preferences.
Moral duty has also been an enduring aspect of the labor movement ethos,
and is still evident as institutions akin to the Catholic Church and the United
Dawkins 217
Nations endorse labor unions as vehicles for improving working conditions
and recognizing human potential (Pope Paul XXIII, 1991; Thomas 2009;
United Nations, 2008), but it also implies ethical behavior by labor unions.
Social contract theory, which is derived from the writings of philosophers
such as Thomas Hobbes, John Locke, and Jean-Jacques Rousseau, details an
implicit moral and ethical agreement between society and an organization
whereby society authorizes the organization in return for its contributions to
important societal goals. Hasnas (1998) states that the social contract is
essentially comprised of social provisions or stipulations that require busi-
nesses to benefit: (a) consumers through economic efficiency; (b) workers
through employment opportunities and workplace dignity; and (c) society by
avoiding practices that degrade a given group or entity (e.g., worker exploita-
tion, environmental ruin). In order to warrant continued authorization, the
advantages an organization provides to society must outweigh any disadvan-
tages it poses (Donaldson, 1982). Because they are economic and social orga-
nizations authorized by society, the provisions of the social contract apply to
labor unions as well. Thus, in addition to the imperative of responsible opera-
tion to maintain legitimacy, and the ethical duty to advance workplace pro-
tections and human potential, LUSR is warranted because it is the means
through which labor unions discharge their social contract obligation to pro-
vide economic, workplace, and social benefits to society.
To whom are labor unions responsible. With regard to the second question,
labor unions are responsible to stakeholders. Freeman (1984, p. 46) defined a
stakeholder as “any group or individual who can affect or is affected by the
achievement of the organization’s objectives.” The essence of stakeholder
theory is that managers should create and sustain moral relationships, and
make good on the affirmative obligation to fairly distribute the harms and
benefits of their organization’s activities (Donaldson & Preston, 1995; Free-
man, 1994).3 According to Evan and Freeman (1993) the obligation to stake-
holders is derived from Kant’s principle of respect for persons, which holds
that persons are entitled to be treated not merely as a means to the achieve-
ment of the ends of others, but as valuable ends in themselves. Thus, organi-
zations are morally obligated to address the interests of their stakeholders and
direct resources and activities to their benefit.
According to Dawkins (2010), labor union stakeholders can be placed in
economic, workplace, and social categories, which correspond to the stipula-
tions of the social contract. Economic stakeholders include (a) union mem-
bers who need competitive wages and benefits, (b) consumers that desire
consistent delivery of products and services, and (c) management and share-
holders who call for economic efficiency. Workplace stakeholders include (a)
218 Business & Society 55(2)
union members who desire equitable treatment, (b) nonunion workers who
constitute an opportunity for union growth and derive a collateral benefit
from labor union initiatives, and (c) managers who desire a stable, orderly
workplace. Finally, social stakeholders include (a) disenfranchised persons
(domestic and international) who desire freedom of association, freedom
from forced and child labor and employment discrimination, and nonexploit-
ative wage levels, (b) workers who desire an adequate social safety net and
safe work conditions, and (c) environmental entities that lack dedicated
representation.
The respect for persons underpinnings of stakeholder theory help leaders
to broaden their view of the organization’s responsibilities to include interests
of less prevalent groups and entities, but provides no clear direction for rec-
onciling the competing interests of stakeholders. While assessing the relative
standing of stakeholder interests remains a considerable challenge for organi-
zational leaders, the LUSR model provides a means with which to systemati-
cally examine those interests and determine a suitable response.
Components of Discharging Labor Union Social
Responsibility
Thus far, it has been argued that labor unions have institutional, ethical, and
social contract obligations to return benefits to the societies within which
they operate, and stakeholder theory has been employed to specify those to
whom labor unions are responsible. As McWilliams and Siegel (2001) have
noted, social responsibility does not necessarily require a radical departure
from organizational activities, but rather an extension of those activities
beyond narrow self-interest. It follows that a socially responsible labor union
will discharge its obligations to society by, not only addressing its members,
but also the economic, workplace, and social interests of their other stake-
holders.
The economic component addresses the traditional “business unionism”
(Dubofsky & Dulles, 1993; Hattum, 1993) role of bargaining for better
wages, benefits, and job security. As the exclusive bargaining agent and stew-
ard of union dues, a labor union’s primary responsibility is to assure that its
members receive an equitable share of the economic rewards they help to
produce. All other labor union roles are premised on this fundamental duty.
Consequently, it is to be expected that the majority of labor union activities
will be directed toward economic equity. The pursuit of economic equity is,
however, bounded by the financial constraints of corporations and the needs
of consumers for consistent delivery of important goods and services.
Dawkins 219
The workplace component entails providing union members with a means
of workplace democracy by which to influence the tenor of their work lives.
However, just as corporations cannot focus solely on shareholders, the focus
of socially responsible labor unions extends beyond the interests of union
members. Workplace democracy also includes the interests of supervisors,
the management and union hierarchies that jointly administer the collective
bargaining agreement, and all workers. Suitable activities include addressing
management needs for flexible scheduling and problem-solving contribu-
tions from workers, prohibitions of child and sweatshop labor, and broad
advocacy for a workplace that is both humane and efficient.
The social component pursues the objective of social justice, particularly
for the community of workers and marginalized sectors of society to whom
labor unions have historically appealed (Cornfield, 1991). Social justice per-
tains to the distribution of benefits and burdens in the economic, political,
and environmental systems and entails involvement in the political process to
influence outcomes in those areas. Lastly, the social component involves
advocacy on issues of broad concern such as the impact of globalization on
human and worker rights around the world. Injustices in the broader social
environment make it less likely that workers and less powerful stakeholders
will achieve economic equity and workplace democracy. The economic,
workplace, and social components are not mutually exclusive or cumulative.
That is, activities in each area constitute only part of LUSR and can be pur-
sued simultaneously. Labor unions in varying degrees have pursued objec-
tives in each of these areas, and continue to do so. This LUSR model serves
to facilitate analysis of labor union activities and call attention to all aspects
of social responsibility.
Labor Union Social Responsibility and Union
Member Attachment
In addition to LUSR’s normative value, the extent to which perception of
LUSR positively impacts key measures of union attachment such as union
commitment, job satisfaction, propensity to participate, and propensity to
withdraw, make LUSR a potentially useful construct. Gordon, Philpot, Burt,
Thompson, and Spiller (1980) describe union commitment as containing four
major constructs; an attitude of loyalty to the union, a feeling of responsibility
to the union, a willingness to exert strong effort on behalf of the union, and
a belief in the goals of unionism. Labor union social responsibility has the
potential to impact union commitment because, although it is related to the
societal impact of union activity, it is likely to parallel pro-union attitudes.
220 Business & Society 55(2)
There is evidence that pro-union attitudes and ideology are foundations for
the development and maintenance of union commitment (Bacharach,
Bamberger, & Sonnenstuhl, 2001; Gordon et al., 1980). Consequently,
Tetrick and Barling (1995) and Newton and Shore (1992) have argued that
unions must make a greater investment in developing prounion attitudes as a
source of union commitment.
Labor union social responsibility also has the potential to influence com-
mitment through perceived external prestige, what relevant others think
about an organization (also called construed external image and organiza-
tional prestige; see Smidts, Pruyn, & Van Reil, 2001). Organizational com-
mitment was influenced by perceived external prestige (Carmeli, 2005), and
socially responsible companies are more attractive to potential workers and
tend to have more committed workers (Brammer, Millington, & Rayton,
2007; Turban & Greening, 1997). There is also evidence that persons like to
be associated with socially responsible voluntary organizations (Boezeman
& Ellemers, 2007). In the same way, union members’ perceptions of unions
improve when unions are diverse (Bacharach & Bamberger, 2004), or engage
in altruistic activities (Fiorito, 1992), and positive perception of unions pre-
cede other measures of union attachment such as union participation (Gordon,
Barling, & Tetrick, 1995).
Hypothesis 1: Labor union social responsibility is positively related to
union commitment.
Labor union social responsibility can enhance job satisfaction by improv-
ing external aspects of job satisfaction and the labor relations climate in the
workplace. Job satisfaction is an evaluative judgment workers make about
their job and is derived from the extrinsic rewards, or the intrinsic tasks and
responsibilities of the job itself (Weiss, 2002). Union members view the
internal aspects of job satisfaction such as task complexity, degree of auton-
omy, and opportunities for promotion less favorably than do nonunion work-
ers (Freeman & Medoff, 1984), and it is difficult for labor unions to influence
these factors. Labor unions can; however, favorably impact extrinsic factors
of job satisfaction such as pay, and work rules and procedures (Bryson,
Cappellari, & Lucifora, 2003; Farber & Western, 2002; Renaud, 2002) and
socially responsible unions will attempt to do so. Socially responsible labor
unions also have the capacity to improve the labor relations climate. A poor
labor relations climate reduces job satisfaction (Artz, 2010), and thus it is
reasonable to infer that a good labor relations climate will increase job
satisfaction.
Dawkins 221
Hypothesis 2: Labor union social responsibility is positively related to
job satisfaction.
According to Barling, Fullagar, Kelloway, & McElvie (1992) there are a
number of variables that predict union commitment, but job satisfaction is
among the most prevalent. How job satisfaction affects union commitment is
not entirely clear. One view proposed by Newton and Shore (1992) and
Iverson and Kuruvilla (1995) is that the impact of job satisfaction on organi-
zational commitment is mediated by other attitudes. Barling et al. (1992)
have argued that job satisfaction has a direct and independent effect on union
commitment. Finally, Bamberger, Kluger, and Suchard (1999) conducted a
meta-analytic study of job satisfaction and union commitment research and
found that models proposing direct and mediated effects of union commit-
ment on job satisfaction fit better than models positing direct or mediated
effects alone. One conclusion that can be drawn from this group of studies is
that job satisfaction regularly mediates the relationship between attitudinal
variables and union commitment.
Hypothesis 3: The positive relationship between labor union social
responsibility and union commitment is mediated by job satisfaction.
Union participation is a behavioral expression of union commitment
(Parks, Gallagher, & Fullagar, 1995) involving activities such as attendance
at union meetings, talking up the union, volunteering time and effort to ben-
efit the union, and voting in union-sponsored elections (Shore & Newton,
1995). Organizational withdrawal entails the intent to reduce job inputs and
work-role inclusion (Hanisch, Hulin, & Roznowski, 1998), but is demon-
strated in the union context by diminished involvement in voting, recruiting,
and other activities, or disassociating with the union altogether.4 Participation
in local union activities has been a consistent and positive consequence of
union commitment (Bamberger et al., 1999; Tan & Aryee, 2002). Moreover,
Fullagar, Clark, Gallagher, & Carroll (2004) have demonstrated that the
impact of union commitment on union participation persists over time. Snape
and Chan (2000) add that commitment precedes participation in union activi-
ties because commitment provides the motivation to participate. Besides
being the inverse of participation, union withdrawal is likely to have a nega-
tive association with union commitment because there is no motivation to
participate.
222 Business & Society 55(2)
Hypothesis 4a: Union commitment is positively related to union
participation.
Hypothesis 4b: Union commitment is negatively related to propensity
for union withdrawal.
As noted by Hammer, Bayazit, and Wazeter (2009), meta-analytic studies
of union commitment and participation show that job satisfaction has direct
or indirect influence that, in turn, contributes to union participation.
Withdrawal indicators such as intent to quit are often explained as conse-
quences of two mediating variables, organizational commitment and job sat-
isfaction, both of which are proposed to have a negative effect on turnover
cognitions (Price, 2001; Somers, 1995). Because job satisfaction coincides
with favorable activity such as organizational citizenship behaviors and is
negatively associated with intentions to quit, it is likely to mediate the rela-
tionship between LUSR and union participation and propensity to withdraw
from the union. Commitment is also likely to mediate the relationship
between LUSR and union participation and propensity to withdraw from the
union because it mediates the impact of attitudes on withdrawal intentions
and cooperative intent (Boezeman & Ellemers, 2007).
Hypothesis 5a: Job satisfaction and union commitment will mediate the
positive relationship between labor union social responsibility and
union participation.
Hypothesis 5b: Job satisfaction and union commitment will mediate
the negative relationship between labor union social responsibility
and propensity to withdraw from the union.
Method
This section explains the methodology of the study: Participants and proce-
dure; measures; and the analytic strategy for structural equation modeling.
Participants and Procedure
The respondents in this study were union carpentry apprentices enrolled in
an apprenticeship-training program jointly sponsored by a craft union and an
association of building contractors under terms of their collective bargaining
agreement. Apprenticeship programs are designed to equip full-time workers
(apprentices) with skills in all aspects of a particular craft through supervised
on-the-job training and related in-class theoretical instruction delivered by
certified union trainers (Bilginsoy, 2007). Craft unions organize workers on
Dawkins 223
the basis of common skill (e.g., electricians, plumbers), whereas industrial
unions tend to organize based on industry (e.g., steel, auto). The apprentices
are craft union apprentices, and will generally complete journey-level certi-
fication (one who has fully learned the trade) in 3 to 5 years.
With the help of union instructors who were blind to the purposes of the
study, two separate surveys were administered representing the independent
and dependent variables respectively to craft union apprentices in the south-
western United States. Participation was voluntary and confidentiality was
promised, but not anonymity, because the two-stage data collection required
respondents to identify themselves such that their surveys could be linked
together at the individual level. The first survey instrument was administered
containing the independent variables (Time 1) prior to the start of quarterly
classroom training and the second survey containing the mediating, and
dependent variables was administered three days later (Time 2).5 At Time 1,
325 respondents completed the survey, and approximately 96% of Time 1
respondents also completed surveys at Time 2 (312 of 325). The respondents
were all men with a mean age of 26 years and mean union tenure of approxi-
mately two years (standard deviation = 0.11). There are no controls for gen-
der, age, or occupation in the study because of the homogeneity of the group.
Measures
There were five variables: The independent variable LUSR; two mediating
variables, job satisfaction and union commitment; and two dependent vari-
ables, union participation and propensity for union withdrawal. All of the
items for the variables were scored using five-point (1 = strongly disagree,
5 = strongly agree) Likert-type scale items with higher values indicating
greater evidence of the underlying construct.6 In addition, language experts
were employed to perform forward and back translations of the survey such
that it could be administered in English and Spanish.7 A three-step process
was used to develop the LUSR scale. First, items were derived by reviewing
research on labor union roles (e.g., Godard, 1997) and posing questions based
on those studies to focus groups of union members and nonunion continuing
education students. Second, prominent corporate social responsibility scales
were reviewed (e.g., KLD Analytics Scale and Boston University Corporate
Citizenship Survey) and 12 items were developed by adapting questions from
those surveys to labor unions. Third, items were pilot tested with continuing
education students not previously involved in the focus groups.
Next factor analysis, a statistical method used to examine how underlying
constructs influence the responses on a measured variable, was employed.
224 Business & Society 55(2)
Confirmatory factor analysis seeks to determine if the number of factors and
the loadings (relation of measured item to underlying construct) conform to
what is expected on the basis of pre-established theory, model, or rationale.
Indicator variables are selected on the basis of prior theory and factor analy-
sis is used to see if they load as predicted on the expected number of factors.
The researcher’s a priori assumption was that each factor (the number and
labels of which may be specified) was associated with a specified subset of
indicator variables. In this case the factors were proposed components of
LUSR. Based on factor analysis of the pilot study results, three subscales of
LUSR were developed reflecting the economic, workplace, and social com-
ponents. The subscales included items such as labor unions improve wages
and benefits, labor unions help find positive solutions to workplace problems,
and labor unions care about consumers (composite reliabilities = 0.78, 0.86,
and 0.81 respectively), and were combined to comprise LUSR.8
Union commitment was assessed with an index derived from the Sverke
and Kuruvilla (1995) scale. The index was comprised of two subscales of
four items each for the value-rational and instrumental-rational components
of union commitment, and included items such as “I feel that I am an impor-
tant part of my union” and “my union’s chances of improving my work situ-
ation are good.” The mean scores of each subscale were combined to represent
union commitment (composite reliability = 0.83 and 0.88 respectively). Job
satisfaction was measured with five-items derived from the Fricko and
Beeher (1992) scale and included items such as I enjoy coming to work each
day and I get a feeling of satisfaction from my work (composite reliability =
0.85). Union participation was measured with four items adapted from the
Shore and Newton (1995) scale and included items such as I am more likely
than other union members to vote in union elections (composite reliability =
0.89). Lastly, propensity for union withdrawal was measured with two items
adapted from the Hom, Griffeth, and Sellaro (1984) intention to quit ques-
tionnaire (composite reliability = 0.75), that included items such as if I could
get a similar job without a union, I would.
Analytic Strategy
In order to estimate models and test hypotheses, structural equation modeling
(SEM) was used with AMOS software (Arbuckle, 2003). Structural equation
modeling essentially combines factor analysis and multiple regression tech-
niques to test the validity of a proposed theoretical model and to analyze the
structural relationships between independent (i.e., exogenous) and depen-
dent (i.e., endogenous) variables (Schumacker & Lomax, 2004) jointly.
Dawkins 225
Structural equation modeling is generally comprised of two steps: validat-
ing the measurement model (i.e., confirmatory factor analysis) and fitting
the structural model (i.e., multiple regression path analysis). The process is
based on a proposed theory or model of relationships, which is tested.
Construct indicators are developed for each variable and verified with con-
firmatory factor analysis. Items can be dropped from the model to improve
fit—the extent to which the data conform to the hypothesized constructs—
after which the proposed theoretical model can be evaluated against a null
or competing model (Garson, 2011).
Like many statistical techniques, SEM has proponents and detractors.
Kelloway (1995) identifies the primary criticisms of SEM as the qualitative
nature of structural equation models whereby theorists assume that a group of
variables are causally related and test propositions about them, the assess-
ment of global fit, the use of specification (adjusting theoretical models to fit
data), and its complexity. Conversely, SEM permits the testing of complete
models rather than individual variables, permits testing of models with mul-
tiple dependent variables, and permits modeling of mediating variables rather
than being restricted to an additive model such as in ordinary least squares
regression. Moreover, whereas regression is highly susceptible to errors of
interpretation by misspecification, the SEM strategy of comparing alternative
models to assess relative model fit is more robust. The general conclusion is
that, appropriately employed, SEM has the potential to substantially advance
the understanding of organizational phenomena (Garson, 2011; Kelloway,
1995). As shown by the hypotheses, this article has a theoretical model, sev-
eral constructs that must be tested jointly, multiple dependent variables, and
mediating variables. Given these characteristics, the author judges that SEM
is an appropriate analytical tool.
Anderson and Gerbing (1988) recommend a two-stage process to con-
ducting SEM whereby a researcher validates the measurement model with
confirmatory factor analysis before completing structural analysis of the
model. If the indicator variables load as predicted on the expected factors
then the confirmatory factor analyses reveals that the model fits the proposed
theoretical model. Goodness of fit tests determine if the model being tested
should be accepted or rejected and reflect the outcome of confirmatory factor
analysis. These overall fit tests do not, however, establish that particular
paths within the model are significant. If the model is accepted, the researcher
will then employ structural analyses to interpret the path coefficients in the
model (Garson, 2011).
A battery of fit statistics is generally employed in SEM analyses. Initially,
the chi-square goodness of fit to degrees of freedom ratio test is employed.
226 Business & Society 55(2)
The chi-square statistic is used to test the hypothesis that the relationships
proposed in the model provide a plausible explanation of the measured data.
The chi-square statistic, however, is sensitive to sample size, and is thus often
coupled with degrees of freedom to provide a more robust measurement. To
further assess the fit of each of the tested models, Schumacker and Lomax
(2004) recommend the following additional tests: (a) the Tucker–Lewis index
(TLI), (b) the comparative fit index (CFI) (c) root-mean-square error of
approximation (RMSEA), and (d) standardized root-mean-square residual
(SRMR). Satisfactory model fit is indicated by ratio of chi-square goodness
of fit to degrees of freedom no greater than two (Browne & Cudeck, 1993),
TLI and CFI values no smaller than 0.90, and RMSEA and SRMR values no
greater than 0.08, and 0.10 respectively (Schumacker & Lomax, 2004). The
values used to assess fit are analogous to significance levels in regression and
other statistical techniques. Additionally, Hu and Bentler (1999) suggest
assessing model fit using a combination of SRMR, CFI, and TLI, and chi-
square difference tests to ensure a robust assessment of the hypothesized
measurement and structural models.
For the confirmatory analysis of the measurement model, all of the indica-
tor items were included for each of the independent variables and revisions
were based on examination of the modification indices, standardized residu-
als, and indicator reliabilities provided in the AMOS output.9 To guard against
multicollinearity, the reliability of the variables was evaluated by calculating
composite reliabilities and average variance extracted as recommended by
Anderson and Gerbing (1988). Composite reliability (Fornell & Larcker,
1981) draws on the factor loadings generated in the confirmatory factor anal-
ysis to produce a measure of internal consistency comparable to coefficient
alpha. Average variance extracted indicates the ratio of total variance that is
due to the latent variable.
To determine the significance of the hypothesized paths and test media-
tion, bootstrapping—a nonparametric approach to hypothesis testing that
does not assume a mathematical distribution, but still permits empirical esti-
mates of the standard errors—was employed (Shrout & Bolger, 2002).
Sampling distributions associated with indirect effects (i.e., mediating vari-
ables) are often non-normal, which compromises the statistical power of tra-
ditional parametric tests. To avoid this difficulty, nonparametric bootstrapping
procedures are recommended when examining indirect effects because
assumptions of normality regarding the underlying sampling distribution are
not required (Bollen & Stine, 1990).
Dawkins 227
Results
Inadequate variance, nonnormal data, and multicollinearity can present
substantial problems with SEM and hence they are addressed here.
Preliminarily, the data were screened to locate ill-scaled items, which have
variances larger than 10 times the smallest variance, as this can cause prob-
lems in SEM (David, Kline, & Yang, 2005) and it was determined that all
of the variances were within the acceptable range. Next, given the detri-
mental impact of nonnormality (kurtosis in particular) on maximum-
likelihood estimates and their standard errors and model fit statistics in
SEM, the observed indicator variables were screened for out-of-range
skewness and kurtosis. Skewness values ranged from −1.40 to 0.98
(Median = 0.69), and kurtosis values ranged from −0.02 to 2.40 (Median =
0.10) falling within suggested ranges of –/+ 3 for skewness and –/+ 10 for
kurtosis (David et al., 2005; Kline, 1998).
Lastly, it was determined whether multicollinearity posed unacceptable
Type II error rates. Grewal, Cote, and Baumgartner (2004) advise that multi-
collinearity can cause substantial Type II error rates in SEM models when:
(a) composite reliability is weak, 0.70 or lower; (b) explained variance is low,
0.25 or below, and (c) sample size is relatively small (e.g., less than a 3:1 ratio
between respondents and the variables tested). However, as composite reli-
ability improves to .75 or higher, explained variance reaches 0.50, and the
sample becomes relatively large (6:1 ratio), Type II error rates are reduced to
below five percent. As shown by the descriptive statistics in Table 1, the com-
posite reliabilities, averages of explained variance, and sample size for this
study comport with the required conditions. Thus, the amount of Type II error
risk posed by multicollinearity is within a statistically acceptable range. Also,
confirmatory factor analysis demonstrated the discriminant validity of the
constructs. Lastly, it is conceptually consistent for the some of the subscales
(e.g., union commitment and LUSR) to be highly correlated because they
represent components of the larger construct.
Measurement Model
As shown in Table 2, based on confirmatory factor analysis, the initial
measurement model, which compares the data with the proposed theoreti-
cal model, was a relatively poor fit. The chi-square to degrees of freedom
statistic was above two and neither of the fit indices exceeded the recom-
mended threshold of .90 (TLI = 0.88, CFI = 0.89). Model re-specification—
eliminating questionnaire items to improve the fit of the data with the
228 Business & Society 55(2)
Table 1. Descriptive Statisticsa,b, Zero Order Correlations, Reliability, and Average
Variance Extracted Estimatesc.
Variable Mc SD 1 2 3 4 5 6 7 8
LUSR–economic 4.13 .75 .62/.66
LUSR–workplace 3.94 .75 .61** .84/.65
LUSR–social 3.77 .83 .60** .72** .81/.69
Value-rational
commitment
3.90 .82 .41** .49** .43** .83/.77
Instrumental-
rational
commitment
3.94 .83 .45** .57** .57** .69** .88/.84
Job satisfaction 4.39 .62 .31** .41** .41** .41** .49** .85/.77
Union
participation
3.43 .97 .26* .41** .47** .45** .44** .36** .89/.85
Propensity to
withdraw
3.14 1.58 –.14** –.23** –.16* –.22** –.26** –.27** –.10* .75/
.68
aScale ranges from 1 to 5. N = 312 for all variables.
bMean scores were computed by summing the final items and dividing by the number of items.
cComposite reliability and Average Variance Extracted estimates are at the end of each row.
LUSR = labor union social responsibility; M = mean; SD = standard deviation.
**Significant at the 0.01 level (2-tailed). *Significant at the 0.05 level (2-tailed).
Table 2. Goodness-of-Fit Results for Measurement Modelsa
Model χ2 df Δ χ2 TLI χ2/df RMSEA CFI SRMR
Revised measurement
model
674.67** 410 .94 1.65 .05 .95 .06
Initial measurement
model
1516.16** 770 841.49** .88 1.93 .06 .89 .06
Independence model 5360.17** 465 4685.50** 11.53 .18
One-factor model 2006.50** 431 1331.83** .65 4.65 .11 .68 .91
Three-factor modelb 1333.45** 428 658.78** .80 3.12 .08 .84 .071
Note. N = 312. Change in chi-square was calculated independently by contrasting the different
models against the hypothesized models. TLI = Tucker–Lewis Index (Tucker and Lewis, 1973);
RMSEA = root-mean-square error of approximation (Steiger, 1990); CFI = comparative fit
index; SRMR = standardized root-mean-square residual.
aThe variables were labor union social responsibility, union identity, instrumental union com-
mitment, value-based union commitment, union activity, and propensity to quit.
bThree factors are LUSR items on a single factor, union commitment and job satisfaction items
on a second factor, and participation and propensity for union withdrawal items on the third
factor.
**p = < .001.
Dawkins 229
proposed theoretical model—is, however, to be expected in confirmatory
factor analysis (Anderson & Gerbing, 1988). Thus, after considering factor
loadings, squared multiple correlations, and modification indices, seven
questionnaire items10 were eliminated that did not correlate well with the
other items and constructs being measured. As shown in the revised mea-
surement model in Table 2, this action resulted in a measurement model
with TLI and CFI fit indices that exceed the recommended threshold of
0.90.
For purposes of comparison, the hypothesized measurement model was
contrasted with the less constrained independence model—representing the
null hypothesis—and two constrained nested models (i.e., submodels) in
which the items were set to load on specific factors. Because the survey was
comprised entirely of self-report measures, the goodness of fit of a single
factor model was also examined to test for common method bias. If a single
factor accounts for the majority of variance in the observed variables, it may
be indicative of common method variance. In this case, the poor fit of the
single-factor measurement model provides evidence against bias from com-
mon method variance (Podsakoff, MacKenzie, Lee, & Podsakoff, 2003).
Lastly, given the high intercorrelations between job satisfaction and union
commitment, a three-factor model was created by loading the LUSR items
on a single factor, the union commitment and job satisfaction items on a
second factor, and the participation and propensity for union withdrawal
items on the third factor. The hypothesized measurement model fit the data
significantly better (TLI = 0.94, CFI = 0.95, RMSEA = 0.05, SMSR = 0.06)
than either of the alternative models (see Table 2), indicating that the items
converged as intended. In addition chi-square goodness of fit to degrees of
freedom (χ2/df = 1.65) was below the 2.0 threshold established by Browne
and Cudek (1993).
Assessing concurrent and discriminant validity provides further support
for the validity of the model. Concurrent validity concerns whether the cen-
tral concepts, such as the specific components of LUSR, actually predict
other constructs that they could be expected to predict; discriminant valid-
ity refers to differences between the three components in terms of divergent
relationships to other variables. Consistent with the model, the results sug-
gest that the three LUSR commitment components correlate with each
other, but are distinct, and demonstrate both discriminant and concurrent
validity.
230 Business & Society 55(2)
Structural Model and Support for Hypotheses
Table 3 displays the standardized factor loadings for the indicators used in
the revised measurement model, and Table 4 displays the fit statistics for the
structural model. The fit statistics indicated good overall fit with the data
(χ2/df = 1.57, TLI = 0.94, CFI = 0.95, RMSEA = 0.04, SRMR = 0.05)
according to the recommended cutoffs. As shown in Figure 1, the hypothe-
sized model explained a considerable amount of variance in union commit-
ment (69%), and moderate amounts of variance in job satisfaction (26%),
union participation (35%), and propensity for union withdrawal (25%).
Table 5 presents the direct and indirect effects and associated confidence
intervals. Hypothesis 1 predicting a positive relationship between LUSR and
union commitment was confirmed with a beta coefficient of 0.63 and signifi-
cant at the .001 level (ß = 0.63, p < .001). The positive beta indicates that
union commitment generally follows LUSR, in the sense that commitment
tends to move up when LUSR moves up. Moreover, the p-value of >.001 indi-
cates that the relationship between LUSR and union commitment was almost
certainly not attributable to chance. Hypothesis 2 predicting a positive rela-
tionship between LUSR and job satisfaction was confirmed as well (ß = 0.51,
p < .001). Hypothesis 3 predicted that the positive relationship between LUSR
and union commitment would be mediated by job satisfaction and was sup-
ported. The indirect effect of LUSR on union commitment was positive and
significant (ß = 0.16, p. < .01). Tests of hypotheses 4a and 4b confirmed the
predicted positive relationship between union commitment and union partici-
pation (ß = 0.59, p < .01) and negative relationship between union commit-
ment and propensity for union withdrawal (ß = −0.33, p < .01) respectively.
Hypothesis 5a predicted an indirect and positive relationship between LUSR
and union participation and was supported (ß = 0.45, p < .01). Finally,
Hypothesis 5b predicting an indirect and negative relationship between LUSR
and propensity for union withdrawal was supported (ß = −0.34, p < .01).
To further assess whether the hypothesized indirect effects of LUSR on
union participation and propensity to quit were fully or partially mediated,
two additional models were tested including distinct direct paths between
LUSR and union participation and propensity for union withdrawal. To com-
plete the analysis the hypothesized model was used as the basis for compari-
son with the nested-models, direct paths between LUSR and union
participation and propensity for union withdrawal were included, and changes
in the fit indices were examined. The significance level of the change in chi-
square between the hypothesized model and the additional models reflect the
effects of the added paths, providing a test of model fit. Table 4 shows the
Dawkins 231
Table 3. Measurement Model Indicator Loadings.
Variable Questionnaire item Loading
LUSR–economic 1 Labor advocate wage demands that
improve the standard of living for
workers.
.89
LUSR–economic 2 Labor unions increase job security. .41
LUSR–workplace 1 Labor unions respond to members’
concerns.
.67
LUSR–workplace 2 Labor unions help find positive solutions
to workplace problems.
.71
LUSR–workplace 3 Labor unions ensure workers health and
safety.
.78
LUSR–workplace 4 Labor unions ensure fair treatment of
workers
.71
LUSR–workplace 5 Labor unions give workers a “voice”
similar to that of management.
.77
LUSR–workplace 6 Labor unions enable workers to have a
say in what happens to them at work.
.66
LUSR–social 1 Labor unions care about consumers.
.70
LUSR–social 2 Labor unions support sound
environmental practices.
.79
LUSR–social 3 Labor unions do volunteer work in the
community.
.68
LUSR–social 4 Labor unions speak up for disadvantaged
people.
.81
LUSR–social 5 Labor unions reduce discrimination in the
workplace.
.74
Job satisfaction 1 I enjoy coming to work each day. .77
Job satisfaction 2 I get a feeling of satisfaction from my
work.
.71
Job satisfaction 3 I am considering leaving carpentry for
another career (reverse scored).
.68
Job satisfaction 4 The job fits well with my lifestyle.
.61
Job satisfaction 5 The longer I am on the job the better the
job gets.
.62
Value-based
commitment 1
I believe in the goals of my union.
.65
Value-based
commitment 2
I feel that I am an important part of my
union.
.73
Value-based
commitment 3
My union means a great deal to be
personally.
.83
(continued)
232 Business & Society 55(2)
results of the models that include the additional direct paths and neither of the
additional direct paths produced a significant change in model fit suggesting
that the added paths were not significant. This outcome suggests that the
effect of LUSR on union participation and propensity for union withdrawal
was fully mediated by job satisfaction and union commitment.
Discussion
This study offers an empirical test of the impact of LUSR on the job satisfac-
tion, union commitment, union participation, and propensity for union
Variable Questionnaire item Loading
Value-based
commitment 4
I feel a strong sense of belonging to my
union.
.71
Instrumental
commitment 1
My union’s chances of giving me more say
in the way I do work are good.
.70
Instrumental
commitment 2
My union’s chances of offering me
employment security are good.
.73
Instrumental
commitment 3
My union’s chances of improving my
physical work environment are good.
.76
Instrumental
commitment 4
My union’s chances of fixing a problem
are good.
.68
Union participation 1 I am more likely than other union
members to vote in union elections.
.86
Union participation 2 I am more likely than other union
members to read the union newsletter.
.88
Union participation 3 I am more likely than other union
members to attend general membership
meetings.
.77
Union participation 4 I am more likely than other union
members to help with union activities.
.61
Propensity to
withdraw 1
All things being equal, I would take
a job that does not require union
membership.
.69
Propensity to
withdraw 2
If I could get a similar job without a union,
I would.
.65
Note. LUSR = labor union social responsibility.
All loadings were significant at p < .01.
Table 3. (continued)
Dawkins 233
Table 4. Goodness-of-Fit Results for the Structural Modelsa.
Model χ2 df Δ χ2 TLI χ2/df RMSEA CFI SRMR
Hypothesized
structural model
704.99** 449 .94 1.57 .04 .95 .05
Independence model 5351.18** 496 10.79 .18
Additional direct path
LUSR–Union
participation
703.41** 448 1.58 .94 1.57 .04 .95 .05
LUSR–Propensity for
union withdrawal
704.98** 448 .01 .94 1.57 .04 .95 .05
Note. LSUR = labor union social responsibility; TLI = Tucker–Lewis Index (Tucker and Lewis,
1973); RMSEA = root-mean-square error of approximation; CFI = comparative fit index;
SRMR = standardized root-mean-square residual.
Economic
(.80)
Workplace
(.94)
Social
(.74)
Labor Union
Social
Responsibility
Job
Satisfaction
(.26)
Union
Commitment
(.69)
Union
Participation
(.35)
Propensity
to
Withdraw
(.25)
Instrumental
(.94)
Value-based
(.70)
.90**
.97**
.86**
.51**
*
.32**
.63***
.59** -.33**
.97**
.83**
Figure 1. Path Model of Labor Union Social Responsibility and Indicators of Union
Engagement.
Arrows are accompanied by path coefficients.
***Significant at .01 level
**Significant at .01 level
*Significant at .05 level
aNumbers in parentheses represent r-squared statistic.
234 Business & Society 55(2)
withdrawal of union members. Labor union social responsibility explained
35% of variance in union commitment and 26% of variance in job satisfac-
tion. As predicted, union commitment and job satisfaction subsequently
mediated the impact of LUSR on union participation and propensity for
union withdrawal. These results suggest that members who perceived labor
unions to be more socially responsible were more committed to the union,
more satisfied with their jobs, more likely to participate in union activities,
and less likely to withdraw from the union. The LUSR construct demon-
strates predictive capacity in that all of the hypotheses predicting direct and
mediated impacts on aspects of union engagement were supported. These
results are the only empirical test of the Dawkins (2010) three-component
model of LUSR and provide preliminary evidence for its validity.
Although the findings invite discussion of causal relations, further testing
is required to more thoroughly examine the links between the components of
LUSR and union engagement. With regard to measurement, the objective was
Table 5. Standardized Direct and Mediated Effects and the Associated 95%
Confidence Intervals.
Direct effects Estimate Lower Upper P
LUSR–Union commitment .63 .50 .74 .00
LUSR–Job satisfaction .51 .36 .65 .00
Job satisfaction–Union
commitment
.32 .20 .43 .00
Union commitment–Union
participation
.59 .48 .68 .00
Union commitment–Propensity to
withdraw
–.33 –.46 –.21 .00
LUSR–Union participation .14 –.19 .48 .31
LUSR–Propensity for union
withdrawal
–.01 –.33 .57 .93
Mediated Effects–Labor union social responsibility
Union commitment (via job
satisfaction)
.16 .09 .24 .00
Union participation (via union
commitment)
.45 .34 .57 .00
Propensity for union withdrawal
(via union commitment)
–.34 –.48 –.21 .00
Note. The upper and lower bounds of the 95% confidence interval (shown in parentheses) were
based on the findings from bootstrapping (e.g., nonparametric hypothesis testing) analysis.
Dawkins 235
to develop a valid measure of LUSR with conceptually distinct components.
The confirmatory factor analysis of the economic, workplace, and social com-
ponents revealed that it is possible to distinguish among three components of
LUSR. The distinction is important because in order for a union to improve its
social responsibility it must be able to identify the areas of its shortcomings.
On the other hand, the components of LUSR are clearly correlated. The cor-
relations between the economic, workplace, and social components are, how-
ever, consistent with previous research on broad-based constructs such as
organizational commitment and union commitment (e.g., Allen & Meyer,
1990; Hackett, Bycio, & Hausdorf, 1992; Sverke & Kuruvilla, 1995).
The LUSR model can be improved by drawing distinctions between the
overall perception of LUSR and its separate components. For example, is a
union that excels on the economic component but largely ignores the social
component socially responsible? A rigorous assessment on LUSR might
require adequate performance on each component as opposed to permitting
excellent performance on one component to offset poor performance on
another. Conversely, reasonable arguments can be made in defense of labor
unions that focus more heavily on the economic and workplace components
during contract years. The Dawkins model could be markedly improved by
providing guidance for labor union leaders on how to reconcile the compo-
nents of LUSR he identifies.
Implications for Research and Practice
Because the LUSR construct has the potential to explain and predict union
attitudes and activity, there are implications for labor relations research and
practice. First, it can potentially inform and broaden discussions of union
revitalization. Snape and Redman (2004) identify three broad models (strate-
gies) of unionism in the labor relations literature and there are LUSR impli-
cations for each. The “service model” (Bamberger et al., 1999), emphasizes
satisfying the economic needs of members with relatively less attention to
ideological considerations. The “organizing model” (e.g., Bacharach et al.,
2001; Frege & Kelly, 2004) highlights active, self-reliant workplace union-
ism through mutual support and solidarity, and focuses primarily on mem-
bers, but extends beyond their economic and workplace needs to their social
aspirations. The covenantal model (e.g., Herman, 1997) goes further in
stressing, not only economic gains and social support for union members, but
the value-based social movement aspects of unions demonstrated through
partnerships with NGOs, and other social and political activism.
236 Business & Society 55(2)
Proponents of the covenantal model have argued that union members who
identify heavily with the union are more likely to work for the union than
members whose commitment is based heavily on economic premiums (e.g.,
Bronfenbrenner & Juravich, 1997; Burchielli, 2006). Similarly, Freeman and
Rogers (1999) contend that a bottom up resurgence is required for union revi-
talization that begins with union member attachment. Based on the Dawkins
formulation, however, socially responsible labor unions can address the foci
of service, organizing, and covenantal models concurrently in ways that com-
port with their moral and social contract obligations. Because LUSR is a
multifaceted construct wherein the individual components can have differen-
tial impact, it would be interesting to examine how the components of LUSR
severally influence perception and behavior. For example, would the social
component of LUSR, which aligns with the covenantal model, have greater
impact on union member engagement than the workplace and social compo-
nents? If Freeman and Rogers are correct, unions wishing to increase mem-
ber attachment will need to buttress the social component of LUSR while
continuing to deliver vital economic rents and workplace protections.
Secondly, the LUSR model provides a basis for comparisons of strategy
and tactics across public and private sector union settings. Public sector
unions operate under a different regulatory regime in the United States and
have much higher membership density than private sector unions. It may be
that the public perceives the work of public sector unions such as firefighters,
civil service workers, and teachers as having greater social value. Research
directed at extending the LUSR model should focus on validating and
employing measures of LUSR with different types of unions, occupations,
and in countries with different regulatory structures. More broadly, it is plau-
sible that in other countries the regulatory apparatus is so different that the
perception and relative impact of the economic, workplace, and social aspects
of LUSR are quite different than in the United States.
Lastly, framing labor union activity in terms of social responsibility could
influence the tenor of discussions regarding labor relations outcomes. Budd
(2004) identifies efficiency, equity, and voice as the key outcomes of the
labor relations system. Efficiency reflects the priorities of managers, inves-
tors, and consumers for affordable goods and services, while equity and voice
reflect union member desires for reasonable apportionment of economic
rents and the capacity to influence the nature of workplace governance. Labor
union social responsibility adds a social component and also provides a
model with which to discuss the balance of these outcomes in labor relations
systems that reflects a broader consideration of all stakeholder interests.
Dawkins 237
Whereas unions in the United States have traditionally been characterized
by their “bread and butter” focus on economic outcomes (Dubofski & Dulles,
1993; Hattam, 1993), which can lead to poor public perceptions, the success
of programs like the Justice for Janitors campaign in Southern California indi-
cate that the public can be supportive of labor initiatives that highlight social
objectives (Erikson, Fisk, Milkman, Mitchell, & Wong, 2002). This research
suggests that union members (i.e., internal stakeholders) may be similar to
external union stakeholders in desiring to see unions as positive forces both in
their lives and in the broader world community. It may be that if labor unions
demonstrate greater concern for the broad economic, workplace, and social
interests of their stakeholders, they will reap the benefit of higher levels of
commitment among their members, and support from the outside community.
Corporations and managers do not perceive more favorably labor unions that
attend carefully to the social component of LUSR. Given the current level of
labor law violations by employers (e.g., Bronfenbrenner, 2001; Mehta &
Theodore, 2005) that level of good faith is not currently in evidence. An
enduring challenge for LUSR is likely that some objections to labor unions are
not due to their activity, but to their mere existence.
Limitations and Conclusion
The findings of this study must be considered in light of its shortcomings.
First, the generalizability of the study is limited because of the demographic
characteristics of the sample, and the job context. Gender, age, job and
union tenure, and occupation all influence union attitudes and behavior
(Bender & Sloane, 1999; Posthuma, 2009; Schur & Kruse, 1992). Thus,
results obtained from craft union apprentices may not translate to journey
certified craft workers or industrial union members. Clearly, replication
studies with samples drawn from other types of union members, such as
white-collar, industrial, and professional workers will be useful in establish-
ing the generalizability of the LUSR construct. Second, this study was
conducted in the United States and additional research is required to estab-
lish the impact of LUSR more broadly. Given the preliminary nature of the
construct and limited empirical research, it is difficult to determine the
extent to which the findings are unique to this sample. Lastly, all data were
collected using self-report measures from a single source, which raises the
possibility that common method variance influenced the findings. Several
steps were taken to reduce this possibility, including collecting the data in
two waves, and applying the one-factor test, which suggested that common
method bias was not an issue.
238 Business & Society 55(2)
In this article a model of LUSR was tested and the fact that LUSR corre-
lated with measures of union engagement in a manner generally consistent
with the labor relations literature should increase confidence in the meaning-
fulness of the construct. Hence, the study indicates that the prediction of vari-
ous attitudes and behaviors (e.g., union commitment and participation) can
be improved by considering the impact of LUSR. These findings may.
Hopefully, this article encourages others to test the specific relations exam-
ined here more broadly and extensively, and advances research on how labor
unions can manage their social responsibility to serve the interests of their
stakeholders and enhance union viability.
Declaration of Conflicting Interests
The author(s) declared no potential conflicts of interest with respect to the research,
authorship, and/or publication of this article.
Funding
The author(s) received no financial support for the research, authorship, and/or pub-
lication of this article.
Notes
1. For an excellent treatment of labor unions advocacy of these objectives see
Labor in America: A History, Chapters 9 to 11, by Dubovsky and Dulles (1993).
2. The National Labor Relations Act (1935) designates exclusive representation
by one union for all members in a bargaining unit, and prohibits direct dealing
(negotiating with individual union members). Thus, provisions of the collective
bargaining agreement negotiated by that union bind all employees in a bargain-
ing unit that has elected a union.
3. There are also descriptive and instrumental (i.e., strategic) strands of stakeholder
theory, but they are intentionally excluded. Simultaneous treatment of the varia-
tions of stakeholder theory leads, in the author’s view, to confusion rather than
clarity.
4. Prevailing labor relations laws and regulations may prohibit workers from termi-
nating union membership.
5. This interval accommodated the preference of the union leaders because appren-
tices have classroom training one week every three months.
6. The reading level of the respondents was relatively low, which necessitated
some breaks with survey convention. Pilot surveys with this group of respondents
indicated considerable confusion with changes in question format. Consequently,
negatively phrased items were not used.
Dawkins 239
7. Chi-square difference tests of respondents using English and Spanish versions of
the questionnaire were insignificant. Survey items are available from the author
upon request.
8. All of the items used in the analysis are provided in Table 3, after confirmatory
factor analyses and other statistical analyses (i.e., fitting of the model) were com-
pleted.
9. Details regarding these measures are available from the author.
10. The dropped items are
How much money, time, or attention do labor unions give to
1 = not at all, 2 = small amount, 3 = moderate amount, 4 = a large amount, and
5 = very great amount
1. safeguarding the environment
2. providing employment opportunities for ethnic and racial minorities
3. supporting education in communities where you do business
4. encouraging and supporting employee volunteering
5. assuring that the company produces accurate financial reports
6. talking with community leaders and groups about their concerns
7. having family friendly work policies contributing to charities.
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Author Biography
Cedric E. Dawkins (PhD, The Ohio State University) is an associate professor of
management at the Dalhousie University. His research interests include labour rights,
corporate disclosure, and stakeholder engagement. His work appears in Business &
Society, Business Ethics Quarterly, and Journal of Business Ethics.
http://www.unglobalcompact.org/aboutthegc/thetenprinciples/index.html
Labor Unions and Support for Redistribution
in an Era of Inequality∗
David Macdonald, Florida State University
Objectives. The United States has become increasingly unequal over the past several decades. Despite
this, public opinion toward redistribution has remained largely unchanged. This is puzzling, given
Americans’ professed concern regarding, and knowledge of, rising inequality. I argue that the decline
of labor unions, an organization that promotes anti-inequality attitudes among its members, can
help us to understand this. Method. I use panel data from the 50 U.S. states from 1978 to 2012
and ordinary least squares regression to examine how state-level unionization levels condition the
relationship between income inequality and support for redistributive spending. Results. I find that in
contexts where labor unions are stronger, higher levels of income inequality prompt greater suppor
t
for welfare spending. Conclusion. These findings illustrate an additional mechanism through which
labor unions can check income inequality and help us to understand why the American public has
not turned in favor of redistribution during an era of rising economic inequality.
The United States has become increasingly unequal. Over the past four decades, a
vast amount of wealth has concentrated at the top of the income distribution. Indeed,
U.S. inequality has reached levels not seen since before the Great Depression (Saez and
Zucman, 2016). High levels of inequality enhance the political influence of the affluent,
while depressing that of the poor (Bartels, 2016; Gilens, 2012; Schlozman, Verba, and
Brady, 2012), something that is troubling for democratic governance and political equality
(Hacker and Pierson, 2010; Page and Gilens, 2017).
The decline of organized labor is often viewed by scholars as a driver of rising inequality
(Ahlquist, 2017; Bartels, 2016; Bucci, 2018; Schlozman, Verba, and Brady, 2012; Volscho
and Kelly, 2012). Typically, labor unions are thought to reduce income inequality through
two mechanisms. The first is by boosting their workers’ wages through collective bargaining
(Freeman and Medoff, 1984; Rosenfeld, 2014), which produces a more equal distribution
of wealth between workers and management. The second is through electoral mobilization,
by bringing a higher share of low- and middle-income voters into the electorate (Leighley
and Nagler, 2007), and increasing the political influence of the less affluent (Flavin, 2016).
These are not the only mechanisms, however, through which labor unions can influence
economic inequality. I argue that there is another mechanism through which unions can
influence inequality. Specifically, I argue that unions promote anti-inequality attitudes
among their members. Where organized labor is stronger, there are more individuals who
hold anti-inequality attitudes and, thus, by extension, the mass public is more responsive to
rising inequality, that is, people turn in favor of redistribution when inequality rises. To my
∗Direct correspondence to David Macdonald, Department of Political Science, Florida State University,
Tallahassee, FL 32306 〈dm14@my.fsu.edu〉. The author will gladly share all data and coding for replica-
tion purposes. All data and replication code, as well as the supplemental appendix, are publicly available on
the Harvard Dataverse.
SOCIAL SCIENCE QUARTERLY, Volume 100, Number 4, June 2019
C© 2019 by the Southwestern Social Science Association
DOI: 10.1111/ssqu.12627
1198 Social Science Quarterly
knowledge, no study to date had investigated the conditioning effect of unionization on the
relationship between inequality and public opinion about government welfare spending.
I draw upon recent work that illustrates labor unions’ capacity to shape their members’
political attitudes and clarify their economic self-interest (Ahlquist and Levi, 2013; Iversen
and Soskice, 2015; Kim and Margalit, 2017; Mosimann and Pontusson, 2017). Using
survey data from the 2012 American National Election Study (ANES), 2004 National An-
nenberg Election Study (NAES), and aggregate panel data across the 50 U.S. states from
1978 through 2012, I show that union members are more aware of, and more concerned
about, rising income inequality. Furthermore, in states with higher levels of union mem-
bership, the mass public responds to rising inequality by demanding more redistribution.
These findings contribute to our understanding of the political consequences of declining
union membership, highlighting one such consequence—decreased support for economic
redistribution. They can also shed light on a theoretically puzzling relationship between
income inequality and support for redistribution, specifically that there has been little
change in public demand for redistribution, even among the less affluent, despite decades
of rising inequality (Kelly and Enns, 2010; Shaw and Gaffey, 2012). This is particularly
puzzling, given that Americans favor lower levels of inequality (Page and Jacobs, 2009;
McCall, 2013), and are well aware that it has risen over the past several decades (Kelly and
Enns, 2010).
I argue that it is not simply political ignorance, apathy regarding inequality, nor ac-
ceptance of an unequal status quo (Bartels, 2016; Hayes, 2014; Trump, 2018) that has
prevented U.S. public opinion from turning in favor of economic redistribution over the
past several decades. The decline of labor unions, an organization that promotes pro-
redistributive attitudes, and conditions public responsiveness to income inequality, has
also played an important role. The decline of unions has political implications, not only
depressing voter turnout among the less affluent (Leighley and Nagler, 2007), but also
weakening a source of pro-redistributive attitudes. This can in turn reduce public pressure
on government to address the issue of high, and rising, income inequality.
Inequality and Public Opinion Toward Redistribution
Models of democracy and inequality (Meltzer and Richard, 1981) assume that when
inequality rises, the public will respond by demanding greater redistribution because the
median voter can be better off economically by supporting government redistribution. In
this sense, democracy should be “self-correcting” in terms of addressing income inequality,
with higher inequality leading to greater demand for redistribution. This has not been
observed in many democracies (Benabou, 2000), including the United States (Kelly and
Enns, 2010; Luttig, 2013; Wright, 2018), which stands out as one of the most unequal
OECD (Organisation for Economic Co-operation and Development) countries.
1
Polls consistently show that Americans are opposed to high levels of inequality (Page and
Jacobs, 2009; McCall, 2013), and that people are well aware that inequality is higher today
than in decades past (Kelly and Enns, 2010).2 Despite this, there has been little observed
responsiveness to rising inequality, that is, public opinion shifting in a pro-redistributive
direction, and we lack a satisfactory answer as to why this is the case.
1Among the 35 OECD countries, the United States ranks as the third most unequal, after post-
tax redistribution, with a Gini coefficient of 0.39. Only Turkey, Chile, and Mexico ranked higher
〈https://data.oecd.org/inequality/income-inequality.htm〉.
2See 〈http://www.aei.org/wp-content/uploads/2015/04/Political-Report-May-2015 〉.
Unions and Support for Redistribution 1199
Not all works find that people are unresponsive to income inequality, with several studies
finding that higher levels of local inequality depress belief in American meritocracy (New-
man, Johnston, and Lown, 2015), and boost support for organized labor (Newman and
Kane, 2017), and political candidates who support inequality-reducing policies (Newman
and Hayes, 2017). Experimental evidence also shows that information regarding the extent
of inequality can prompt support for redistribution (Boudreau and MacKenzie, 2018;
McCall et al., 2017). Overall, however, findings are mixed (Franko, 2016), and in the
aggregate, we have observed little movement in support for redistribution during an era of
rising income inequality. I argue that the decline of labor unions can help us to understand
why a mass public that cares about inequality and is aware of its rise has not responded in
the expected direction—by demanding greater government redistribution.
How Labor Unions Influence Opposition to Inequality
There are theoretical reasons to expect a link between union membership and attitudes
toward inequality. The primary objective of labor unions is to bargain on behalf of workers,
primarily seeking to increase wages, and promote a more egalitarian distribution of earnings
in the workplace by “compressing pay,” that is, increasing the wages of rank-and-file
employees and limiting the compensation of management (Freeman and Medoff, 1984).
Indeed, unionized workers have higher wages and more generous benefits than do their
nonunionized counterparts (Rosenfeld, 2014; Western and Rosenfeld, 2011), and working
poverty is lower in U.S. counties with high levels of unionization (Brady, Baker, and
Finnigan, 2013). Labor unions have been linked to more liberal public policy and lower
inequality (Becher, Stegmueller, and Käppner, 2018; Bucci, 2018; Franko, Kelly, and
Witko, 2016; Kelly and Witko, 2012; Radcliff and Saiz, 1998), as well as greater demand
for redistribution and liberal public policy (Franko, 2016), at the U.S.-state level, and
greater political equality in representation (Ellis, 2013; Flavin, 2016). In a recent review,
Ahlquist (2017) notes that higher levels of unionization have been linked to lower levels of
income inequality, both in the United States and cross-nationally.
Politically, labor unions have long favored a higher minimum wage, universal health-
care, more generous pensions, and a strong social safety net. Unions have also protested
against high levels of executive compensation, particularly at the expense of rank-and-file
employees (Lichtenstein, 2013; Western and Rosenfeld, 2011). Furthermore, unions dis-
proportionately support the pro-redistributive (compared to the Republicans) Democratic
Party (Anzia and Moe, 2016; Dark, 1999; Jansa and Hoyman, 2018), endorsing candidates
and mobilizing their members for political action on behalf of union-backed candidates
and ballot initiatives (Asher et al., 2001; Flavin and Radcliff, 2011; Flavin and Hartney,
2015; Francia and Bigelow, 2010; Francia and Orr, 2014; Kerrissey and Schofer, 2013;
Radcliff and Davis, 2000; Zullo, 2004) . The egalitarian norms that labor unions promote,
along with general support for left-leaning politicians and liberal policies, should not be lost
upon their members. Indeed, a recent literature has examined the ability of labor unions to
influence their members’ political attitudes, arguing that unions disseminate information
and communicate with members, and through their political activity and collective bar-
gaining tactics, promote norms of egalitarianism and altruism (Ahlquist, 2017; Ahlquist
and Levi, 2013).
In short, I argue that labor unions promote economically egalitarian attitudes among
their members and increase opposition to economic inequality. When organized labor is
1200 Social Science Quarterly
FIGURE 1
Union Communications to Members on the Issue of Income Inequality
stronger, that is, in a particular state, that means there are more individuals in the mass public
who are union members, and are thus being exposed to an “anti-inequality” environment.
Illustrating the Mechanisms
I argue that labor unions shape their members’ attitudes on the issue of income inequality,
shifting them in a more egalitarian direction, through two mechanisms: direct information
provision and facilitation of workplace discussion. Though I cannot directly test these
mechanisms, that is, observe specific union communications to their members on the
issue of inequality, and then assess how union members internalized this information, nor
directly observe union-based discussion about economic inequality, I attempt to bring a
variety of indirect evidence to bear, showing that, first, labor unions do emphasize the
issue of inequality and, second, that unions—by facilitating workplace discussion—can
shape their members’ attitudes on this issue. To make this argument, I draw on recent
work that has demonstrated the capacity of labor unions to shape their members’ political
attitudes on economic issues (Ahlquist, Clayton, and Levi, 2014; Kim and Margalit, 2017;
Mosimann and Pontusson, 2017).
Labor unions provide their members with information about political issues, via newslet-
ters, emails, and social media. See, for example, Figure 1, which shows a recent tweet and
report on pay differentials from the AFL-CIO, the largest union organization in the United
States, emphasizing the norm of wage equality and economic fairness, informing members
about the issue of income inequality, and clearly emphasizing organized labor’s position.
The AFL-CIO not only sends out communications to its members, but also publishes a
yearly report on pay differentials, entitled the “Executive PayWatch Database.” These data
are prominently featured on the AFL-CIO’s website, and list the highest-paid CEOs in the
United States and show the ratio between the CEO and the median employee for hundreds
of companies, from Walmart—where the CEO makes over $9,000 an hour compared to
$9.00 an hour—to McDonald’s, Johnson & Johnson, and Capital One.3 This informa-
tion is highlighted by union leadership such as AFL-CIO President Richard Trumka, and
3See 〈https://pr-paywatch-aflcio.pantheonsite.io/paywatch/company-pay-ratios〉.
Unions and Support for Redistribution 1201
TABLE 1
Union Membership and Workplace Discussion of Politics
0 Days 1–2 Days 3–4 Days 5–7 Days
Union members 31.2% (1,969) 24.9% (1,542) 19.5% (1,210) 23.9% (1,483)
Nonunion members 38.1% (18,183) 26.4% (12,585) 17.4% (8,278) 18.0% (8,600)
NOTE: The sample is restricted to people who are currently employed. The number of observations is in
parentheses. A bivariate regression of the number of days discussing politics in the past week (0–7) on
union membership (1,0) was statistically significant (t = 13.03, p = 0.000).
SOURCE: 2004 NAES.
receives attention and coverage from the mass media and from prominent politicians.4
For example, in a 2017 speech, AFL-CIO President Richard Trumka said that “income
inequality was tearing apart the United States and the entire world” and “the mission of
the labor movement is to fight back against the forces responsible for widening the income
gap.”5
In addition, union organizations such as these send members information on salient
legislation, via an online legislative scorecard.6 This provides information on how legislators
voted and illustrates the union’s stances on key issues such as the 2018 Fiscal Year Budget
Resolution. The AFL-CIO opposed this bill, arguing that “it cuts programs like Medicare,
Medicaid and other income security programs that provide critical support to our nation’s
most vulnerable population by $4.1 trillion over the next decade to pay for trillions of
dollars in tax giveaways for millionaires and major corporations.”
These are certainly not the only ways through which unions can provide information to
their members, but they are illustrative of one such way—via online activity.
Another is through workplace discussion. Data from the 2004 NAES, presented in
Table 1, show that union members are significantly more likely to discuss politics at work,
a finding consistent with Kerrissey and Schofer (2013), who show that union members
are significantly more politically active and engaged than their nonunion counterparts.
Although I cannot directly observe workplace discussion about inequality, it seems plausible
that at least some political discussion centers around the issue of economic inequality,
particularly given that union leadership frequently emphasizes issues relating to wage
equality and leveling of the economic playing field.
Information dissemination such as this, as well as workplace discussion of political
issues, has the potential to influence members’ political attitudes. Indeed, past research
has found that union members pay attention to the policy stances taken by their union
and form their attitudes based in part on the position of their union, following not only
top-down cues from the union, but also being influenced by workplace discussion and
union-based socialization.
Ahlquist, Clayton, and Levi (2014) illustrate this in a study of dockworkers on the
U.S. West Coast in the International Longshore Warehouse Union (ILWU). They found
that members of the union were more willing to take a protectionist stance on trade
policy, following the ILWU’s position, even though this stance cut against their material
4See 〈https://aflcio.org/press/releases/ceo-pay-increases-347-times-average-workers〉; 〈http://money.cnn.
com/2018/05/22/news/economy/ceo-pay-afl-cio/index.html〉; 〈http://www.aei.org/publication/hillary-and-
bernie-both-complain-about-excessive-ceo-pay-but-the-average-ceo-makes-less-than-hillarys-speaking-fee/〉.
5See 〈https://www.peoplesworld.org/article/as-afl-cio-convention-opens-trumka-emphasizes-need-to-
battle-income-inequality/〉.
6See 〈https://aflcio.org/scorecard/votes〉.
1202 Social Science Quarterly
TABLE 2
Union Membership and Knowledge of Rising Inequality
Much Somewhat About Somewhat Much
Larger Larger the Same Smaller Smaller
Union members 67.1% (378) 17.9% (101) 11.5% (65) 1.6% (9) 1.8% (10)
Nonunion members 56.3% (2,948) 22.1% (1,159) 15.7% (824) 3.9% (202) 1.9% (102)
NOTE: Question asks about the difference in incomes between rich and poor people in the United States
today compared to 20 years ago. The number of observations is in parentheses. A bivariate regression
of knowledge of rising inequality (1–5) on union membership (1,0) was statistically significant (t = 4.62,
p = 0.000).
SOURCE: 2012 ANES.
self-interest. Ahlquist, Clayton, and Levi also found that ILWU members who had more
exposure to union communications and who engaged in more frequent workplace dis-
cussion were more likely to have an opinion on trade policy, and were more likely to
have an opinion that is consistent with the union position. Specifically, ILWU members
were more likely to support restrictions on imports and less likely to express confidence in
NAFTA. Ahlquist, Clayton, and Levi attribute these findings to ILWU communications via
the union newspaper, workplace discussion, and a “socialization” mechanism that shapes
workers’ political attitudes.
Kim and Margalit (2017) similarly found that union positions on trade policy influenced
members’ attitudes, showing that workers (across a variety of industries) were more aware
of their union’s position on trade policy—particularly when union communications were
more frequent. Workers were also more likely to hold a trade policy position in line with
their union compared to nonunionized workers in the same industry. Kim and Margalit
also showed that a sharp shift in the United Auto Worker’s (UAW) union in 2010, from
supporting to opposing the U.S.–Korea Free Trade Deal, led to a substantively large and
significant shift in UAW members’ attitudes on this policy, a shift consistent with changes
in the UAW’s position. There was no similar shift among nonunionized workers in the
auto industry. Kim and Margalit attribute this to the ability of union communications,
workplace discussion, and socialization to influence people’s political attitudes.
Clearly, labor unions have the ability to shape their members’ attitudes on trade policy.
Mosimann and Pontusson (2017) use European Social Survey data from 2002 to 2014 to
show that this extends to redistributive attitudes as well. They argue that this results from
unions’ promotion of egalitarian norms and facilitation of workplace discussion that allows
for information to spread (Iversen and Soskice, 2015; Putnam, 2000), as well as a sense
of altruism and “linked fate” with other workers (Ahlquist and Levi, 2013) that unions
promote. This, Mosimann and Pontusson show, makes all workers, but particularly those
in unions with a larger share of low-income employees, an environment where compression
of wage differentials is more likely to be pursued, and thus norms of egalitarianism more
likely to be emphasized, more favorably inclined toward reducing income inequality.
In short, both top-down union information dissemination and workplace discussion of
politics, which can help inform and socialize union members, should shape attitudes toward
economic inequality. If this was not the case, then we should not observe any differences
between union and nonunion members’ attitudes on the issue of economic inequality.
Table 2 shows that union members are significantly more likely than their nonunion
counterparts to be aware of rising income inequality, with 67 percent of union members
stating that the differences between rich and poor have gotten much larger in the past
Unions and Support for Redistribution 1203
TABLE 3
Union Membership and Attitudes Toward Rising Inequality
Bad Thing Neither Good nor Bad Good Thing
Union members 62.5% (331) 28.9% (153) 8.7% (46)
Nonunion members 51.5% (2,553) 38.2% (1,893) 10.3% (512)
NOTE: In 1967, households in the top 20 percent earned an average of 11 times as much as households
in the bottom 20 percent. Today, the top earn an average of 15 times as much. Is it good, bad, or neither
good nor bad that the DIFFERENCE between the top and the bottom incomes has changed this way? The
number of observations is in parentheses. A bivariate regression of attitudes toward rising inequality (1–3)
on union membership (1,0) was statistically significant (t = 4.13, p = 0.000).
SOURCE: 2012 ANES.
TABLE 4
Union Membership and Inequality Attitudes by Right-to-Work Status
(1) (2)
Knowledge of Inequality Aversion to Inequality
Union member 0.225∗∗∗ 0.197∗∗∗ 0.181∗∗∗ 0.162∗∗
(0.064) (0.064) (0.067) (0.067)
Right-to-work state −0.123∗∗∗ −0.112∗∗∗ −0.183∗∗∗ −0.178∗∗∗
(0.032) (0.032) (0.033) (0.033)
Union member × RTW 0.075 0.103 0.081 0.101
(0.129) (0.129) (0.129) (0.130)
Republican – −0.490∗∗∗ – −0.312∗∗∗
– (0.031) – (0.031)
Constant cut1 −2.104∗∗∗ −2.304∗∗∗ −1.336∗∗∗ −1.453∗∗∗
(0.042) (0.046) (0.028) (0.033)
Constant cut2 −1.627∗∗∗ −1.833∗∗∗ −0.121∗∗∗ −0.229∗∗∗
(0.031) (0.037) (0.023) (0.026)
Constant cut3 −0.841∗∗∗ −1.033∗∗∗ – –
(0.024) (0.028) – –
Constant cut4 −0.214∗∗∗ −0.386∗∗∗ – –
(0.022) (0.025) – –
Observations 5,798 5,782 5,488 5,469
NOTE: Dependent variables (DVs) range from 1 to 5 (column 1) and from 1 to 3 (column 2); Ordered probit
coefficients; robust standard errors are in parentheses.
∗∗∗p< 0.01; ∗∗p < 0.05; ∗p < 0.1 ; two-tailed.
SOURCE: 2012 ANES.
20 years compared to 56 percent of nonunion members. Table 3 provides further evidence
to suggest that income inequality looms larger in the minds of union members, showing
that union members are significantly more likely than their nonunion counterparts to state
that the rise of inequality in the United States over the past four decades is a bad thing,
with 63 percent of union members volunteering this opinion compared to 52 percent of
nonunion members.
Addressing Self-Selection into Labor Unions. Table 4 shows that this relationship
between union membership and attitudes toward inequality holds when taking partisanship
(Bartels, 2016) into account, and that there are also no significant differences in right-to-
work (RTW) and non-RTW states. The RTW analyses are meant to help rule out the
possibility that self-selection is driving therelationship between union membership and
1204 Social Science Quarterly
TABLE 5
Descriptive Statistics, U.S. State Panel, 1978–2012
Obs. Variable Mean SD Min Max
1,750 Pro-welfare spending (%) 20.326 4.843 10.000 39.70
0
1,750 Top 1 percent income share 14.091 4.553 4.600 36.100
1,750 Union membership (%) 14.850 7.089 2.300 38.300
1,450 Private-sector union membership (%) 9.275 4.930 1.100 26.100
1,450 Public-sector union membership (%) 33.452 17.135 5.200 73.100
1,712 Partisanship (Democrat (%)–Republican (%)) 6.663 16.209 −80.052 58.938
1,712 Ideology (liberal (%)–conservative (%)) −14.781 12.572 −64.624 53.813
1,750 Nonwhite population (%) 20.357 14.232 0.500 76.800
1,750 Per capita income (thousands of dollars) 36.576 7.664 20.951 68.957
1,750 Unemployment rate 6.021 2.103 2.300 17.800
1,750 Economic policy liberalism (higher = liberal) 0.009 1.032 −2.100 3.131
attitudes toward inequality. In RTW states, people cannot be forced to join a labor union
as a condition of employment. As such, workers in RTW states can “free ride” and receive
union benefits even if they do not belong to the union. Workers in RTW states have fewer
incentives to join the union, and if they do belong are more likely to have joined as a
result of political predispositions, rather than economic incentives.7 If self-selection was
driving results, that is, political predispositions are driving the decision to join a union and
shaping attitudes toward inequality, we would expect that union membership would not
have an influence on attitudes toward inequality in non-RTW states, where people should
be less likely to join based on political predispositions. Results in Table 4 show that there
are no differences between union members in RTW and non-RTW states, suggesting that
unionization in and of itself is associated with greater knowledge of, and aversion toward,
income inequality.
Overall, labor unions promote a more egalitarian distribution of income, and support
political candidates and policies that favor a strong social safety net, mobilizing their
members for political action in support of these policies. This is reflected in the observed
differences between union and nonunion members’ attitudes toward income inequality.
As such, in contexts where organized labor is stronger, we should observe that the public
is more averse to high inequality, and thus more likely to respond to rising inequality by
supporting government policies to reduce it. More specifically, in contexts where union
membership is higher, public opinion should respond to rising inequality by demanding
increased redistribution.
Hypothesis: Unionization levels condition the relationship between income inequality and
public support for redistribution.
Data and Methods
To test this hypothesis, I employ an ordinary least squares regression model and panel
data from the 50 American states from 1978 to 2012. The three primary variables of
interest are over time measures of income inequality, union membership levels, and public
7See Kim and Margalit (2017) for an extended discussion of this logic and an additional applied example.
Unions and Support for Redistribution 1205
support for welfare spending. Table 5 shows descriptive statistics for each of these three
variables as well as for several controls. I explain each of these in greater detail below.
Dependent Variable
I use a state-level measure of support for government spending on welfare. These data
are obtained from Kim and Urpelainen (2017). Welfare spending is a prominent form of
redistributive spending, comprises a large proportion of state budgets, and has long been
an outcome of interest in the state politics literature (Barrilleaux, Holbrook, and Langer,
2002; Brown, 1995; Key, 1949). Kim and Urpelainen (2017) used multilevel regression
and poststratification (MRP) and data from the General Social Survey to construct over
time measures of state-level support for several different types of government spending.8
This measure is valuable because of longitudinal availability, encompassing a time period in
which unions were strong and inequality was low, as well as the contemporary era of weak
unions and high inequality.9 The dependent variable is operationalized as the percentage
of a state’s population in a given year that says we are currently spending “too little” on
welfare. I use this measure from 1978 through 2012, spanning much of the era of rising
U.S. income inequality.
Independent Variables
The two main independent variables are a measure of the percentage of the nonagricul-
tural workforce that belongs to a labor union and a measure of income inequality. Data
on state-level union membership are obtained from Hirsch, MacPherson, and Vroman
(2001), who used Current Population Survey (CPS) data to calculate annual measures
of state union membership.10 I use this variable from 1977 through 2011 (temporally
preceding the dependent variable by one year).
Data on state-level income inequality are obtained from Mark Frank.11 I measure in-
equality as the wealth share of the top 1 percent in each state. I do this because the nature
of U.S. income inequality over the past several decades has been a collection of wealth
at the top of the income distribution (Bartels, 2016; Volscho and Kelly, 2012). The top
1 percent reflects popular discourse, that is, “the 99 percent versus the 1 percent,” and
follows the trend of a collection of wealth at the very top of the income distribution, rather
than a collection at the 20th percentile and above, for example.12 This measure of inequal-
ity is based on IRS tax returns, rather than survey data from the U.S. Census Bureau. As
such, it does not “top code” high incomes (above $250,000, for example) and thus provides
8See Warshaw and Rodden (2012) for a broader discussion of MRP’s validity as a technique for developing
state-level estimates of public opinion.
9In contrast, data from Pacheco (2014), which uses a similar methodology to measure state-level public
opinion on welfare spending, only ranges from 1978 to 2000.
10This variable includes both public- and private- sector union membership. I also separately run analyses
for private- and public-sector unions, using available data from 1983 to 2011 (preceding the DV by one year).
Results show that both private- and public-sector unionization levels condition the inequality–redistribution
relationship in the American states. See the supplementary appendix for these models.
11See 〈http://www.shsu.edu/eco_mwf/inequality.html〉.
12In contrast, the Gini coefficient (even if it is based on IRS tax return data), a measure of how equally
income is distributed in a society (ranging from perfect equality, where all households hold the same amount
of income, to perfect inequality, where one household owns all income), is not adequately sensitive to changes
at the top of the income distribution (Franko, 2016:963, note 4).
1206 Social Science Quarterly
a more accurate measure of the true extent of income inequality. This measure is available
from 1917 through 2015, I use it here from 1977 through 2011, preceding the dependent
variable by one year.
I also take into account several factors that past research has linked to redistributive
preferences (Franko, 2016; Franko, Kelly, and Witko, 2016; Kam and Nam, 2008). I control
for state partisanship and ideology (Erikson, Wright, and McIver, 1993) , expecting that
more liberal and Democratic mass publics will desire greater levels of welfare spending.13
I also control for the state unemployment rate, per capita income, the percentage of
the population that is nonwhite, and previous government policy liberalism, accounting
for the thermostatic nature of public opinion (Pacheco, 2013; Wlezien, 1995).14 I also
include year fixed effects, which account for factors such as the party of the president, and
any national-level factors that may influence union membership and income inequality.
I include state fixed effects, which restrict variation of welfare support to within states,
rather than both within and across states. These state dummy variables also account for
time-invariant factors such as state political culture or region, that is, whether a state is a
part of the South or not.
Unions and Public Opinion Toward Income Inequality
Table 6 and Figure 2 illustrate the relationship between income inequality, union strength,
and support for welfare spending in the American states. Results show that unions pow-
erfully condition the inequality–redistribution relationship. From 1978 to 2012, overall
(private and public combined) state union membership ranges from 2.3 percent (South
Carolina in 2006) to 38.3 (West Virginia in 1982). At this lowest level of union mem-
bership, results show that an increase in state inequality (measured by the income share
of the top 1 percent) would actually decrease support for welfare spending, with a one
point increase in the top 1 percent wealth share depressing support for spending more
on welfare by 0.16. Results from the marginal effects plot (Figure 2) show that higher
inequality does not significantly increase support for welfare spending until state union
membership exceeds 13 percent—which is above the 2017 national average of 10.7 percent.
At the highest level of observed union membership (38 percent), a one point increase in
the top 1 percent wealth share leads to a 0.67 increase in support for welfare spending,
a substantively significant effect size, given that the dependent variable (the percentage of
people in a particular state-year that say we are currently spending “too little” on welfare)
ranges from 10.0 to 39.7 percent.15
13Partisanship and ideology are measured as the percentage of the state’s population that self-identifies as
Democrats minus the percentage that identifies as Republican, and the percentage of the state’s population
that self-identifies as ideologically liberal minus the percentage that identifies as conservative, respectively.
14Data on the state unemployment rate, nonwhite population, partisanship, and ideology were obtained
from the Correlates of State Policy Data Set (Jordan and Grossmann, 2017); see 〈http://ippsr.msu.edu/public-
policy/correlates-state-policy〉. Data on state partisanship and ideology are missing for Alaska and Hawaii
from 1977 to 1996. This is why there are not 1,750 observations (50 states × 35 years). Data on
per capita income were obtained from 〈https://www.bea.gov/regional/〉 and adjusted for inflation us-
ing the CPI inflation calculator; see 〈https://www.bls.gov/data/inflation_calculator.htm〉. Data on state
economic policy liberalism were obtained from Caughey and Warshaw (2017); see 〈https://dataverse.
harvard.edu/dataset.xhtml?persistentId=doi:10.7910/DVN/K3QWZW〉.
15In the supplementary appendix, I used alternative measures of inequality. I used data from Mark Frank’s
website; a Gini coefficient that is based on IRS tax returns. This is a pretax measure of income inequality in
the states. I also used a posttransfer (after federal and state benefits are included) measure of inequality that
is based on data from the U.S. Census Bureau. These were originally collected by Franko, Kelly, and Witko
Unions and Support for Redistribution 1207
TABLE 6
Union Membership, Inequality, and Support for Welfare Spending (1978–2012)
Top 1 percent income share −0.205∗∗∗
(0.073)
Union membership (%) −0.283∗∗∗
(0.077)
Top 1 percent × Union membership (%) 0.023∗∗∗
(0.006)
Partisanship (Democrat–Republican) 0.018∗∗∗
(0.005)
Ideology (liberal–conservative) −0.005
(0.006)
Nonwhite population (%) 0.060∗∗
(0.025)
Per capita income (in thousands) 0.091∗
(0.051)
Unemployment rate 0.004
(0.069)
Economic policy liberalism −0.087
(0.344)
Constant 17.542∗∗∗
(2.044)
Year fixed effects? Yes
State fixed effects? Yes
Observations 1,712
R2 0.771
NOTE: DV is the percentage saying we are spending “too little” on welfare. All independent variables
measured at t − 1; ordinary least squares coefficients. Robust standard errors are clustered by state
in parentheses.
∗∗∗p < 0.01; ∗∗p < 0.05; ∗p < 0.1; two-tailed.
Exploring Heterogeneity in the Relationship
Public Versus Private Sector. It is possible that issues of inequality are dis-
cussed/emphasized more in private-sector unions, given that workers tend to be less ed-
ucated, and thus the higher share of lower wage/hourly workers might be focused more
on issues of wage inequality than are more highly educated public-sector workers, for
example, teachers.16 To assess this possibility, I separately ran analyses for private- (mem-
bership ranges from 1.1 to 26.1 percent) and public-sector (membership ranges from 5.2 to
73.1 percent) state-level union membership (including the same controls as in Table 5).
These data are available from 1984 onward. Results, presented in Table 7, show that both
types of unionization positively condition support for redistribution.17 See the supplemen-
tary appendix for the full regression models. The coefficient for private-sector unions is
(2016), ranging from 1976 to 2006. They were extended up through 2014 by Bucci (2018). I use them here
through 2011. Regardless of which measure is used, the top 1 percent (based on IRS data), pretransfer Gini
(based on IRS data), or a posttransfer Gini (based on Census data), I find that state-level unionization rates
significantly condition the relationship between inequality and public support for welfare.
16See 〈https://gspp.berkeley.edu/research/featured/public-sector-employees-are-highly-educated-
underpaid〉.
17This is not to say that certain unions would differentially influence their members’ attitudes, that is,
teachers’ unions on education spending or a steelworkers’ union on free trade. However, on the broad
issue of economic inequality, results here suggest that belonging to a union (vs. not belonging) shapes
redistributive attitudes.
1208 Social Science Quarterly
FIGURE 2
Labor Union Strength, Inequality, and Support for Welfare Spending (1978–2012)
–
.5
0
.5
1
C
on
di
tio
na
l C
oe
ffi
ci
en
t
0 10 20 30 40
State Union Membership %
Marginal Effect of Inequality on Support for Redistribution
NOTE: Based on the ordinary least squares model in Table 6. Dashed lines represent 95 percent confidence
intervals.
larger than for public-sector ones, suggesting that private-sector unions (which are larger
in terms of raw numbers of people) make a larger difference (in terms of promoting mass
responsiveness to inequality) than do those in the public sector, although public-sector
unions matter as well. Given that private-sector unions have declined far more than their
public-sector counterparts, this is a particularly important finding.
South Versus Non-South. I also examine potential differences between the South and
the non-South (using the Census Bureau’s divisions). The South has a larger minority
population, and many southern whites are hostile toward economic redistribution. Indeed,
the racial history of this region could certainly shape residents’ attitudes toward issues of
inequality and economic redistribution (Key, 1949; Soss, Fording, and Schram, 2008).
The main analysis in Table 6 includes state fixed effects, which account for any effect that
state political history, that is, being a part of the regional South, may exert on mass support
for welfare spending.
In Table 8, I split the data into southern and nonsouthern states, separately examining
how unionization conditions the relationship between inequality and support for welfare
spending. In both South and non-South, I find that unionization significantly conditions
mass responsiveness to inequality. I find that the relationship is actually stronger in the
South than in the non-South (as indicated by the coefficient on the interaction term). This
suggests that higher unionization would make more of a difference in the generally less
union-friendly and less economically liberal South.
Unions and Support for Redistribution 1209
TABLE 7
Unionization by Sector, Inequality, and Support for Welfare Spending (1984–2012)
(1) (2)
Private Sector Public Sector
Top 1 percent income share −0.156∗ −0.201∗∗
(0.087) (0.083)
Private-sector union (%) −0.413∗∗∗ –
(0.143) –
Top 1 percent × Private-sector union (%) 0.035∗∗∗ –
(0.010) –
Public-sector union (%) – −0.140∗∗∗
– (0.039)
Top 1 percent × Public-sector union (%) – 0.008∗∗∗
– (0.002)
Constant 24.217∗∗∗ 28.075∗∗∗
(2.333) (2.791)
Controls? Yes Yes
Year fixed effects? Yes Yes
State fixed effects? Yes Yes
Observations 1,424 1,424
R2 0.735 0.737
NOTE: DV is the percentage saying we are spending “too little” on welfare. All independent variables are
measured at t − 1. Robust standard errors are clustered by state in parentheses.
∗∗∗p < 0.01; ∗∗p < 0.05; ∗p < 0.1; two-tailed.
TABLE 8
Unionization, Inequality, and Support for Welfare Spending by Region (1978–2012)
(1) (2)
South Non-South
Top 1 percent income share −0.685∗∗∗ −0.064
(0.176) (0.093)
Union membership (%) −0.675∗∗∗ −0.110∗∗
(0.181) (0.050)
Top 1 percent × Union membership (%) 0.063∗∗∗ 0.009∗∗
(0.015) (0.004)
Constant 19.484∗∗∗ 11.756∗∗∗
(3.274) (1.487)
Controls? Yes Yes
Year fixed effects? Yes Yes
State fixed effects? No No
Observations 560 1,152
R2 0.698 0.767
NOTE: DV is the percentage saying we are spending “too little” on welfare. All independent variables
measured at t − 1. Robust standard errors clustered by state in parentheses.
∗∗∗p < 0.01; ∗∗p < 0.05; ∗p < 0.1; two-tailed.
Union Decline and Prospects for Redistribution in an Era of Inequality
Overall, the totality of evidence suggests that labor unions, via a socialization and
information provision mechanism, play an important role in determining the extent to
1210 Social Science Quarterly
which the mass public turns in favor of redistribution when inequality rises, that is,
whether public opinion is responsive to inequality. These are substantively meaningful
results that have important political implications, as shifts in public opinion can, over time,
influence government policy outcomes (Caughey and Warshaw, 2017; Erikson, MacKuen,
and Stimson, 2002; Soroka and Wlezien, 2010), shaping the distribution of economic and
political power in society.
U.S. labor union membership peaked in 1953, with one in three workers belonging to a
union (Goldfield and Bromsen, 2013). Since that time, unions have declined dramatically.
Today, barely one in ten workers belong to a union. Furthermore, RTW legislation is
on the rise, with several states, including Michigan and Wisconsin, former bastions of
organized labor, passing legislation that curtails the ability of unions to organize workers.
Recent Supreme Court rulings that prevent public-sector unions from collecting fees
from nonunion members will likely further undermine organized labor.18 Elite and media
depiction of labor unions as corrupt, greedy, and undeserving of their benefits, particularly
those in the public sector, can also undermine mass support for organized labor (Kane and
Newman, 2017), something that will further increase economic inequality.
Decreased union strength can depress the political voice of lower-income groups (Schloz-
man, Verba, and Brady, 2012) that differ from the upper class in terms of many policy
preferences, particularly those dealing with redistribution (Gilens, 2009; Leighley and
Nagler, 2013). This further skews policy outcomes in favor of the wealthy (Bartels, 2016;
Gilens, 2012), something that runs counter to the principle of political equality (Dahl,
2006; Lijphart, 1997) that underlies democratic governance. Decreased turnout is not the
only manner through which declining union membership can influence inequality. The
results here suggest that diminished unions will drive down support for redistribution as a
result of fewer people being exposed to an environment that promotes pro-redistributive
attitudes, and make it less likely that the mass public will respond to rising inequality by
demanding greater government redistribution. If the mass public is not putting political
pressure on government, be it through higher turnout, or through shifts in public opinion,
then politicians will have little incentive to pursue redistributive policies.
One limitation of this study is that only welfare spending was examined. Indeed, this has
a pejorative connotation (Gilens, 1999), and it is possible that unions would more strongly
promote responsiveness to other types of redistributive spending such as education (see,
e.g., Franko, 2016). Furthermore, support for policies such as higher taxes on the rich or a
higher minimum wage were not examined. This would be useful to examine going forward.
There are several other potential paths for future work as well. I think it is worthwhile to
examine how the strength of organized labor shapes public opinion in the American states.
For example, it would be interesting to compare mass politics in states with a strong union
history versus those that have typically been hostile to unions, perhaps following the model
of Andrew Gelman and his colleagues in their study of the politics in rich and poor states
(Gelman et al., 2009). Future work would also do well to examine whether rising inequality
is driven more by public-sector or private-sector unions, given that the latter have declined
far more than the former. Finally, future research would do well to examine if income
inequality equally depresses the political participation of union and nonunion members,
or if unions acts as a bulwark against the demobilizing effects of high economic inequality.
It is also crucial to examine whether high inequality spurs unions to lobby government to
reduce inequality, and whether state governments actually respond with policy changes.
18See 〈https://www.nytimes.com/2018/06/27/us/politics/supreme-court-unions-organized-labor.html〉.
Unions and Support for Redistribution 1211
Overall, these findings illustrate how unions shape public opinion toward rising in-
equality. Although the American mass public appears to be aware of and concerned about
rising inequality, public opinion has not meaningfully shifted in favor of redistribution,
even in an era of high, and rising, inequality. The decline of organized labor can offer an
explanation for this. The continued decline of U.S. union membership will help to ensure
that “the heavenly chorus sings with an upper class accent” (Schattschneider, 1960:35),
likely resulting in a continuation of the status quo of economic and political inequality.
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Analyses of Social Issues and Public Policy, Vol. 14, No. 1, 2014, pp. 261–280
This paper is part of an ASAP special collection on the Social Psychology of the Great Recession
Social Class Myopia: The Case of Psychology
and Labor Unions
Bernice Lott*
University of Rhode Island
This article explores the potential for a research agenda that includes scholarship
on working class issues and organized labor. Such an agenda is consistent with the
official mission of American Psychological Association—to advance knowledge
that benefits society and improves people’s lives. I focus on our paucity of interest
in the institution that gives the American working class a voice—the labor union.
We know that work is one of the central focuses in the lives of most people and that
the work experience is deeply implicated in satisfaction with life. The efforts of
organized labor to achieve economic fairness and justice, and a healthy workplace
environment, are intertwined with multiple corollary consequences that constitute
a wide and complex spectrum—from physical job safety and economic security on
one end, to the psychological benefits of heightened self-esteem, respect, dignity,
empowerment, and affiliation on the other—all related to satisfaction with life. In
addition, by advancing and protecting the rights of workers, unions are part of
the larger movement for civil rights.
As noted by Williams (2012, p. 39), “Americans display a convenient tone
deafness when it comes to class.” Both myopia and Williams’ metaphor are ap-
plicable to American psychology. With our discipline’s historical emphasis on the
individual, we seem to have left theoretical and empirical concerns with social
class largely to sociology. Yet, the assumption of classlessness (or that class is ir-
relevant to our concerns) is contradicted by our own data. Class-related measures,
such as education, income, occupation, and access to resources, are significantly
associated with health and welfare. Acknowledgment of the significance of these
indices for people’s lives, families, and communities can be found in our literature
∗Correspondence concerning this article should be addressed to Bernice Lott, Psychology Depart-
ment, Chafee Building, University of Rhode Island, Kingston, RI 02881. Fax: 401-874-2157 [e-mail:
blott@uri.edu].
I am grateful to Steven Mellor for his constructive suggestions.
261
DOI: 10.1111/asap.12029 C© 2013 The Society for the Psychological Study of Social Issues
262 Lott
(see Lott, 2010, 2012; Lott & Bullock, 2001). Yet, understanding and studying
them under the construct of “social class” remains problematic. Especially prob-
lematic has been our discipline’s distancing from issues relevant to the working
class (see Lott, 2002). Different ways to conceptualize and measure social class are
important issues—beyond the focus of this article—but its reality and significance
are supported by a vast empirical literature (Diemer et al., 2012).
We need to reconcile our neglect of social class with the mission of the
American Psychological Association (APA)—“to advance the creation, commu-
nication, and application of psychological knowledge to benefit society and im-
prove people’s lives”—and with our “Core Values” of “Social justice, diversity,
and inclusion” (APA, 2013a). Our objectives engender pride in our organization
and discipline, but our research and applications have been focused primarily
on human beings who, while slowly acknowledged to differ in gender, ethnicity,
and sexuality, continue to be observed with little recognition of socioeconomic
status—thus maintaining a traditional view of diversity.
My focus here is specifically on the paucity of interest in, and neglect of, the
American labor union—despite its continuously active engagement in efforts to
directly and indirectly benefit society and improve people’s lives. These efforts,
while centered on the achievement of economic benefits and a fair and healthy
workplace environment, are deeply intertwined with multiple corollary positive
consequences. The achievements of organized labor constitute a wide and complex
spectrum, from physical job safety and economic security on one end, to the
psychological benefits of heightened self-esteem, respect, and affiliation, on the
other. As concluded succinctly by Pacek (2009, p. 240), “organized labor exerts a
largely positive effect on human well-being.”
Given that work is a major focus in the lives of most people (Coshow &
Radcliff, 2009), lack of attention to efforts by unions to improve and strengthen
the workplace is a serious omission. As noted by Dutt and Radcliff (2009, p. 15),
“the work experience affects not only one’s economic well-being, but one’s sense
of dignity and self-respect” and “the labor union, as the advocate and agent of the
worker in obtaining both income and dignity, has an obvious potential for affecting
an individual’s satisfaction with life.”
Ignoring the Role of Unions
Psychology is not alone in ignoring the role that the labor movement plays
in the lives of American workers. The U.S. mass media tend to not present fair
and extensive discussions of class, and are generally negative in their treatment
of working-class and labor issues. When unions are presented, they are often
described as not negotiating in good faith, with greedy and self-destructive workers
not interested in the larger public’s well-being (Heider, 2004; Martin, 2004).
Despite the fact that skilled and unskilled workers constitute about half of the
Psychology and Labor Unions 263
labor force in the United States, they are seldom featured positively in television
images, stories, news, and entertainment. Martin (2004, p. 11) notes that “Perhaps
the most important characteristic of mainstream coverage of labor is that the media
are often not covering labor at all.” In 2002, there were fewer than five “labor beat”
reporters left at U.S. daily newspapers. A study of 20th century films (Bodner,
2003) found that, while the Hollywood screen portrayed a variety of working-
class characters, it “almost never celebrated the labor radical or the power of
the militant union” (p. 89). And Coshow & Radcliff (2009, p. 294) write about
the “political ambivalence afforded labor movements” that is reflected within the
social sciences.
There are many examples of scholarship in our discipline that is incomplete—
lacking in full attention to relevant issues—by virtue of neglecting to even mention
organized labor. Illustrative is an issue of the American Psychologist that featured
several papers on retirement. Shultz and Wang (2011), as well as Adams and
Rau (2011), discuss the role of Social Security and employer pensions in making
retirement possible for vast numbers of Americans. They also call attention to
the negative influence on plans for retirement of downsizing and reduced pension
and medical benefits. Yet, there is no discussion of the historical and current
relationship between such benefits and labor union membership. Wang, Henkins,
and van Solenge (2011, p. 209) note that “predictors of retirement adjustment
quality . . . are directly associated with different types of resources that retirees
have”—again, there is no mention of unions. Similarly, an APA co-sponsored
conference on work, stress, and health, as described and summarized by Dingfelder
(2011), did not include the role of organized labor as a topic for discussion.
Dobson and Schnall (2009) note the significant disconnect between our
knowledge of the influence of working conditions on mental health and our lack
of attention to organized efforts to improve these conditions. A book on multidis-
ciplinary perspectives of sex discrimination in the workplace (Crosby, Stockdale,
& Ropp, 2007) does not cite in the index any references to either labor unions
or collective bargaining, ignoring, for example, research on sexual harassment in
which union members have served as respondents (cf. Bulger, 2001). An APA
volume on psychology’s contributions to a healthy world (Rozensky, Johnson,
Goodheart, & Hammond, 2004) includes three chapters on the importance of the
workplace, but there is no mention of organized labor. Even a nonmainstream
association like Psychologists for Social Responsibility (2011) overlooked labor
unions in an otherwise inclusive presentation of an action agenda for the promotion
of psychological health and well-being.
One might expect that organized labor would be a prominent topic within the
field of industrial-organizational (I/O) psychology, but that is not the case. I/O
psychology has ignored unions, despite the stated mission of the Society for I/O
Psychology (SIOP) “to ‘promote human welfare through the various applications
of psychology to all types of organizations’” (Zickar, 2004, p. 146). Zickar notes
264 Lott
that, in the 1990s, labor unions were mentioned in the title or abstract of only 1.3%
of published articles in the top 10 most I/O relevant and prestigious journals, and
only four of 3,468 SIOP members in 2003 identified union issues as an area of
interest.
I/O psychology continues to be closely identified with the issues and concerns
of management, a focus that has been evident since the 1930s when federal
legislation sought to curtail many of the overt strategies that companies had used
to fight unionization efforts. Companies began to recruit applied psychologists who
used personality tests to screen out likely union members among job applicants
who showed signs of “neurotic tendencies,” and to administer attitude and morale
surveys to existing employees to identify workplaces that might be susceptible to
a unionization campaign (Zickar, 2004).
Ross Stagner and Arthur Kornhauser were early exceptions in recognizing
the positive values of unions, with Kornhauser emphasizing the well-being of
workers and potential links between workplace conditions and mental health
(Vinchur & Koppes, 2011; Zickar, 2003). A committee of The Society for the
Psychological Study of Social Issues (SPSSI) on “Trade Union Affiliation,” and a
vote by the SPSSI membership in 1938 to unionize its members (Finison, 1986),
also represent early positive recognition of unions. This interest lessened in the
1940s when SPSSI became increasingly involved with World War II and issues
of national morale as well as with debates over the appropriate relations between
social action and science. There was a brief time in the 1980s when I/O psychology
considered the merits of a relationship between organized labor and psychology,
illustrated by Division 17’s Public Policy and Social Issues Committee, formed
in 1978 (cf. Gordon & Nuriek, 1981). Rosen and Stagner (1980) called for I/O
graduate programs to add union-related and union-informed classes, and Huszco,
Wiggins, and Carrie (1984) proposed areas in which psychologists could offer
their services to unions (e.g., drug abuse programs).
Contemporary neglect of labor unions within I/O psychology and an over-
whelming focus on management issues is apparent in the three-volume (2,300-
page) handbook of industrial psychology recently published by APA (Zedeck,
2011). The index has one footnote and two single-page mentions of labor unions
(and no mention of social class). The handbook contains descriptions of research
that may well have profited from the inclusion of union issues and membership.
For example, Griffin and Clarke (2011) studied ways to reduce stress at work and
increase well-being, but they considered only organizational interventions—for
example, management-sponsored health promotion programs and stress coun-
seling. The authors explicitly reject changes in “the work environment,” pay
increases or benefits, as “not always . . . practicable or desirable.” In another
example, Hammer and Zimmerman (2011) write about quality of work life and
discuss many work-family topics and public policies without any mention of
labor unions. Greenberg (2011) explores the important subject of fairness in the
Psychology and Labor Unions 265
workplace, and discusses perceptions of justice, different kinds of justice, and how
to measure organizational justice. We can perhaps guess at the author’s attitudes
toward unions when we read the following: “negative reactions of employees who
believe they have been victims of injustice” can be illustrated by “the ubiquitous
image of placards proclaiming ‘unfair’ hoisted in the air by disgruntled strikers on
picket lines” (p. 271). While noting that labor unrest is likely to stem from feel-
ings of injustice by workers and that “perceptions of injustice . . . take a toll on
employees’ physical and mental well-being” (p. 312), the author does not mention
the potential positive role of organized labor.
A Very Brief Look at U.S. Labor History
In the 1860s, following the abolition of slavery, workers in the United States
began to organize for an 8-hour workday to allow more time to be with their
families, for education, self-improvement, and for leisure. The movement for
shorter hours was often greeted by employers (aided in some cases by state
militias) with broken promises, police harassment and beatings, raids, strike-
breaking, lockouts, and execution of labor leaders convicted of being anarchists.
In 1884, shortly after President Grover Cleveland declared the first Monday in
September to be Labor Day, and a national holiday, he sent 12,000 troops to break
up the Pullman Strike (Flanders, 2012).
The mid-1880s saw the rise of the American Federation of Labor (AFL),
which led the labor movement until the 1930s when the Congress of Industrial
Organizations (CIO) was formed. Skilled workers had found a home in the AFL,
but immigrant workers who were largely unskilled had turned to the Industrial
Workers of the World, which had a radical vision of class struggle. With passage
of the Fair Labor Standards Act in 1938, the 8-hour day became a matter of
federal law. Both the 8-hour day and the 40-hour workweek are the result of
union initiatives and struggles for fair rewards for labor based on belief in a social
contract of reciprocity between employer and worker. The laws that were proposed
and finally enacted were meant “to allow for physical and mental recovery time
from work” (Dobson & Schnall, 2009, p. 113).
New Deal legislation such as the 1935 National Labor Relations Act (the
Wagner Act) gave workers a legal right to choose union membership and engage
in collective bargaining; and established unemployment compensation, overtime
pay, and the minimum wage. This legislation recognized that “collective strength
was essential . . . [in order] for the worker to approach the employer as an equal
. . . [and receive] the dignity of fair treatment” (Zweig, 2000, p. 121). During the
1930s more than 4 million workers joined unions. Nearly 30% of U.S. workers
belonged to unions in 1945, and 35% in the mid-1950s. In 1955, the two major
unions joined together to form the AFL-CIO and, in the 1960s, more than 30%
of the workforce was unionized (Edelman, 2012). According to some analysts
266 Lott
(Kahlenberg & Marvit, 2012; Zweig, 2004), it was organized labor that helped
build a sturdy middle-class in the United States after World War II. During the first
30 years after the war, a strong labor union voice was successful in raising wages
and working standards for both union and nonunion workers (Feeney, 2012).
By 1983, however, only 20.1% of workers were in unions. According to
Martin (2004, p. 32), “labor’s postwar decline” in membership and economic
and political power can be attributed to the “overlapping factors of unfavorable
legislation, employer resistance, labor’s own growing conservatism, and changes in
popular attitudes.” Edelman (2012) suggests the importance, as well, of structural
changes in the economy and a new political vigor by big business. Now, in the
second decade of the 21st century, union membership is at its lowest point—11.3%
in 2012 (35.9% for public sector workers and 6.6% in the private sector) (Bureau
of Labor Statistics, 2013). While 14.4 million workers report union membership,
15.9 million are covered by union contract benefits.
Some new initiatives appear to be revitalizing the labor movement. Living-
wage campaigns begun in 1994, for example, have spread, succeeding in raising
wages and helping to maintain union jobs in cities and states that have adopted
living-wage ordinances (Luce, 2005). Some local rank-and-file direct actions by
warehouse and food workers have also been successful, for example, warehouse
employee strikes at Walmart, and protests in seafood processing and other food-
industry plants.
Anti-Union Tactics and Consequences
“[O]rchestrated resistance to union formation and collective bargaining . . . is
part of [our] . . . country’s history” (Zandy, 2004, p. 43). From the beginning, union
growth was fought through intimidation and harassment of organizers, violence,
arrests and deportations, and by restrictive laws to make organizing and union
membership more difficult (Martin, 2004). Especially damaging to unions were
two federal laws—the Taft-Hartley Act of 1947, which outlawed slowdowns,
wildcat strikes, and sitdowns, and the Landrum-Griffin Act of 1959, which limited
the right to picket during strikes, and required unions to open their records to
federal investigators.
Unions, particularly those in the public sector, are currently targeted for
blame for state budget deficits and fiscal problems, said to result from pensions
and benefits and overpaid city and state workers (Greenhouse, 2011; Moberg,
2011). Anti-union legislation passed by several Republican controlled legislatures
has singled out unions representing municipal workers, police and firefighters,
and especially teachers (Greenhouse, 2012a). In the film “Won’t Back Down,”
a critic for the New York Times (Scott, 2012) noted that teachers union leaders
are portrayed as cynical and self-serving; one teacher as “a lazy, tenured lump of
pedagogical indifference”; and another teacher (one of the two film heroines) as a
Psychology and Labor Unions 267
“once-proud teacher whose idealism and dedication have withered after years of
frustration” (par. 4).
Current employer tactics in the private sector include the “lockout” of union-
ized workers and the hiring of replacements. These were used by the American
Crystal Sugar Company of Minnesota, whose CEO compared the union to a can-
cerous tumor that had to be removed (Diamond, 2011). Greenhouse (2008) has
described the tactics that corporations employ in their fight against organizing,
and their efforts to weaken the labor movement—workers fear asking for raises
and protesting poor working conditions, the pink slips of layoffs, and retaliation
for pro-union activities.
A lucrative “union-avoidance industry” of 2,000 consultants ranges from
“shady outfits . . . to some of the nation’s most reputable law firms” (Greenhouse,
2008, p. 249); these consultants “specialize in keeping the workplace ‘union free’”
(Zweig, 2000, p. 146). Two studies have documented the use of such consultants
by 75% of companies that were facing drives to organize. Recommended by the
consultants are: requiring employees to attend group or one-on-one meetings to
hear anti-union messages from managers; and threatening to close companies
if unions win an election (cf. Greenhouse, 2008). Mellor and Kathi (2011) cite
a study of 62 elections in 2002 that found anti-union tactics used by 98% of
employers, advised by external consultants 82% of the time—with tactics that
included the promise of higher wages and threats to close or relocate the business.
In their study of more than 1,000 nonunion workers, Mellor and Kathi found a
negative relationship between willingness to work for union representation and
fear of reprisals for disclosing an interest in unions.
So-called Right to Work legislation, currently law in 22 states, makes it illegal
for labor contracts to require workers to pay union dues. Workers in these states
earn less, and are less likely to receive employer-sponsored health care (King,
2012); in 2012, six such states were among the 10 in the nation with the highest
rates of unemployment (Rosenthal, 2012). While it is illegal to fire someone for
trying to organize a union, companies do this repeatedly since the penalties—back
pay for the fired employee—are weak and poorly enforced (Kahlenberg & Marvit,
2012). Employers are allowed to appeal and thus delay a union vote.
Simmons and Harding (2009), noting such anti-union strategies as job ex-
porting and an increased use of part-time workers, argue that government and
corporations have both worked against organized labor by decreasing industry
regulations relevant to wages, hours, and the right to organize. The D.C. Circuit
Court, for example, recently ruled that the free speech of employers is violated
when a company is required to post a notice on bulletin boards that federal law
gives workers the right to form a union (Tritch, 2013). In discussing a 2012
Supreme Court ruling (on a union special assessment), Linda Greenhouse (2013,
par. 8) refers to it as “the latest in years of Supreme Court losses for organized
labor on central questions of collective bargaining.”
268 Lott
Table 1. Contributions of U.S. Labor Movement to Individual and Family Health and Welfare
Attention to safety of the workplace
Benefits, pensions, sick leave, and medical insurance
Child labor laws
Fair wages, job security
Minimum wage and overtime pay legislation
Respect on the job
Right to engage in collective bargaining
The 8-hour day and 40-hour workweek
Unemployment compensation
A number of analysts have argued that the media are typically unfriendly
to organized labor, often treating unions “as advocates of arcane work rules,
protectors of inefficient public employees and obstacles to . . . economic growth”
(Dionne, 2011, par. 4). Dionne argues that investors are presented in the media
as heroes, while workers are the sideshows, whose activities, jobs, and lives are
rarely covered, reflecting a “cool indifference to the heroism of those who go
to work every day.” Zweig (2000, p. 56), a union historian, has documented the
“hostility to organized labor” found in the mass media, concluding that, in the
media, “The working class is truly the silenced majority.” Over the past decades,
union membership has been greatest when the media have reported favorably and
positively about unions, and when the government has played a neutral role.
Why Should Psychologists Care?
An APA survey by the newly established Center for Organizational Excellence
(APA, 2013b) found that more than one-third of the American workers in their
sample experience work stress, and less than half felt adequately compensated
or recognized. Another APA survey found that 65% of adults cited work as a
“significant source of stress.” Clearly then, psychologists should be very interested
in the ongoing work of organized labor because of its historical and continued
significant contributions to worker health and welfare. A list of these contributions
is presented in Table 1.
All the goals of traditional union organizing and activism are focused on
contributing positively to the welfare of individuals and families In 2012, for
example, the median take-home pay for union workers was $943 a week, compared
with $742 for nonunion workers (Bureau of Labor Statistics, 2013). That union
workers may work in different industries than nonunion workers is an important
factor, but the wage differential is significant. “It is well documented,” notes Zweig
(2005, p. 103) “that unions and collective bargaining improve the wages, benefits,
and working conditions of their members compared with similar workers not in
Psychology and Labor Unions 269
unions. Union protection for workers also improves their productivity [and] the
quality of the products they produce.”
Organized labor is committed to obtaining and maintaining both economic
benefits and fair and decent workplaces. Beyond the benefits of higher wages, the
union movement in the United States has won what we all now take for granted:
weekends, sick days and medical leave, maternity leave, overtime, worker’s com-
pensation benefits, child labor laws, employer based health coverage, retirement
plans and pensions, vacations, and rules for safe working conditions and practices.
These achievements benefit nonunion as well as union workers, and many are cod-
ified into federal or state law. In summing up the varied positive consequences
of collective bargaining by unions in the United States, Cobble (2007) includes:
increasing the supply of good jobs at all levels of education; increasing political
participation; and “ensuring that individuals have dignity and time for themselves,
their families, their friends, and their communities” (p. 8).
An absolute priority of organized labor is to make the workplace as safe
as possible (Bulger, 2001), and unions have saved lives and prevented serious
injuries. In the coal industry, for example, one study found a decline of at least
one-third in such statistics in mines that are unionized (cf. Meister, 2011); the
major mine disasters in the past few decades have taken place in nonunion mines.
Prior to 1966, employers were not required to, and argued that they need not,
discuss issues of worker health and safety with the union. This changed when
the NLRB ruled that these were legitimate issues for collective bargaining. As
noted by Kelloway (2004, p. 251), unions press for “three basic rights of workers:
the right to know (about hazards in the workplace), the right to participate (in
removing hazards and improving workplace safety), and the right to refuse unsafe
work.” The tragic explosion of a fertilizer plant in West, Texas, took place in a
nonunion plant that hadn’t been federally inspected for 28 years (Eskow, 2013b),
something that a union would not have allowed.
Western and Rosenfeld (2011) argue that unions contribute to the “moral
economy” by reinforcing the value of fairness and placing that value within an
institutional framework. Zweig (2004) makes the point that unions are not just
about money but about respect and fairness—securing an equitable share of what
labor produces and providing “a decent life for the worker and his or her family”
(p. 9). By advancing and protecting the rights of workers, unions are part of the
larger movement for civil rights, and part of the progress made to reduce ethnic and
gender inequities in the economy (Chang, 2003; Isaac & Christiansen, 2002). One
study (Drago, Colbeck, Hollenshead, & Sullivan, 2008) found a positive effect of
faculty unionization on family-related organizational policies. Using faculties in
English and Chemistry as respondents, the investigators found (in the more than
500 colleges and universities that they surveyed) that the unions had promoted
work-family policies that led to a decrease in the use of “bias avoidance” strategies,
that is, behaviors that minimize the appearance of family commitments in order
270 Lott
to enhance career success. Such strategies were found significantly less often in
colleges with faculty unions.
As the active and legal voice of organized workers, unions make significant
and direct contributions to the enhancement of members’ well-being (Gordon,
Jaurequi, & Schnall, 2009). These include increased opportunities for compan-
ionship, unified action to achieve positive goals, social connectedness, a sense of
community, trust, and ability to redress grievances (Mann, 2011; Pacek, 2009).
Union membership provides the opportunity for control and power over some
aspects of one’s work life and family life, thus promoting self-esteem and psycho-
logical well-being and autonomy (Rosenberg, 2009). A sense of empowerment
follows from the experience of having a clear group “voice” and from collective
action (Isaac & Christiansen, 2002; Sharp, 2011).
Radcliff (2005) calls attention to the role unions play in enhancing mem-
bers’ communication, social and problem-solving skills, and in promoting more
knowledgeable and active citizenship. Thus, for example, union members are more
likely than nonmembers to vote in presidential and congressional elections, re-
gardless of income, education, and occupation (Leighley & Nagler, 2007). States
with stronger unions have greater voter turnout. In the 2012 presidential election,
organized labor played a crucial role (Greenhouse, 2012b; Metzgar, 2012), with
almost 20% of voters, nationally, coming from union households.
Fuller and Hester (2001) surveyed members of a steelworkers local in the
southeast and found that union membership can satisfy socio-emotional needs—
“the need for approval, self-esteem, affiliation, and respect” (p. 1096). Similarly, in
a sample of 17 industrialized democracies, union members were found to express
greater satisfaction with their lives (cf. Radcliff, 2005). Persons in countries with
greater levels of unionization showed more evidence of subjective well-being.
There is a documented positive relationship between union membership and
job satisfaction (Artz, 2010), and “job satisfaction is one of the most important
determinants of overall life satisfaction” (Coshow & Radcliff, 2009, p. 286). Job
satisfaction is most likely to follow from job security, a good work environment,
and from having a voice in the conditions of one’s work. Unions encourage social
support networks, connections, and a sense of solidarity, and our psychological
literature provides empirical support for the conclusion that social connectedness
is positively related to subjective well-being.
Coshow and Radcliff (2009) tested the hypothesis that union membership
contributes to quality of life, using survey data from more than 40,000 respondents
in the 48 continental U. S. states, polled annually from 1983 to 1999. They
found strong support, related to union membership density within states. Their
data reinforce empirical findings about the negative mental and physical health
consequences of job stressors from unfair workloads, low levels of on-the-job
autonomy, and being under-rewarded (Dobson & Schnall, 2009), the prevention
of which are all union objectives.
Psychology and Labor Unions 271
A decline in unionized jobs has been correlated with growth in employment
that is insecure, where layoffs can occur, and that does not provide fair pay,
pensions, or health benefits (Kalleberg, Reskin, & Hudson, 2000). Such “bad
jobs” have been shown to negatively impact not just earnings, but also health
and welfare (Raymo et al., 2011). The retail and service jobs available to young
people, where 68% of those between the ages of 16 and 24 are employed, and
where unions are rare, are particularly characterized by high stress from repetitive
tasks, uncertain hours and scheduling, few benefits, limited learning opportunities,
and extremely high turnover (Tannock, 2004).
Inclusion and Diversity
In expanding outreach for membership, unions have become more and more
welcoming to women, immigrants and people of color (Levi, 2001) and have
made a focus on issues of discrimination and family life clear and explicit. New
bargaining concerns, beyond livable wages, safety, health insurance, and retire-
ment benefits, are such family supports as schedule flexibility, eldercare services,
child and after-school care, paid family leave, and part-time work with benefits
(Firestein & Dones, 2007).
In 2005, women were 43% of all workers covered by union contracts, with al-
most 7 million women members, making “organized labor . . . the largest women’s
movement in the country” (Cobble, 2007, p. 6). Unionized women earn an average
of 31% more than women who are not in unions—with education, training, and
education controlled (Bronfenbrenner, 2005). In 2004, they earned an average of
$19.18 per hour compared with $15.05 an hour for women not in unions. In addi-
tion, unionized women are much more likely to have health insurance, pensions,
job protection for pregnant workers, family leave, and job flexibility (Milkman,
2007).
Union organizing victories have been greatest since the 1980s in work sectors
with high percentages of women, and particularly high percentages of women
of color. These include the public sector, health care, home care, hotels, food
services, light manufacturing, support staff in schools, and graduate students and
adjunct faculty in universities (Bronfenbrenner, 2005; Milkman, 2007). One of the
strongest voices in the contemporary union movement is that of National Nurses
United, part of the Service Employees International Union (SEIU) (Sharp, 2011).
And women now hold two of the top three leadership positions in the AFL-CIO,
while several national unions have women presidents.
As noted by Cranford (2007, p. 410), “unions may become sites where women
are able to renegotiate gender inequalities in significant ways.” Thus, for example,
in a study of women janitors in union leadership positions, she found that they
were experiencing both personal and political transformations, empowerment,
increased confidence to assert their views and to challenge the assumption that
272 Lott
unions are the domain of men. A survey mailed to a large sample of rank and
file members in a Northeastern union local (Mellor, Barnes-Farrell, & Stanton,
1999) found that a relationship between the perceived effectiveness of the union
and participation (meeting attendance and offices held) was strongest for ethnic
minority women, “arguably at the highest risk of unfair treatment and outcomes in
the workplace” (p. 342), and next strongest for other women. Unions are expressing
concern about sexual harassment in the workplace, and some have put in place clear
procedures to assist women in reporting abuses and discrimination (Bulger, 2001).
Results from a sample of union women suggest greater willingness to participate
in union activities when their union is perceived as having a low tolerance for
sexual harassment, as indicated by procedures and policies (Mellor & Golay,
2014).
The CIO was an early force in working for civil rights and ethnic equal-
ity, pressing for integrated workplaces, integrated unions, and federal anti-
discrimination legislation. During the famous sanitation workers strike in Memphis
in 1968, to which Dr. Martin Luther King had traveled to lend his support (and
where he was murdered), workers carried signs reading “I Am A Man” (Isaac
& Christiansen, 2002). African American workers now make up a dispropor-
tionately large portion of public sector workers, and 13.4% of all Black workers
are members of unions—the largest proportion of any ethnic group (Bureau of
Labor Statistics, 2013). Unionized Black workers earn 35% more, and unionized
Latino/a workers earn 51% more, than their counterparts who are not in unions.
It is thus not surprising to find that non-Whites in the United States express more
pro-union sentiments than Whites and “see the labor movement as an essential
institution through which they can address their economic, social and political is-
sues” (Chang, 2003, p. 192). There is a current focus on training Latino/a workers,
2.1 million of whom are unionized, and in supporting amnesty and citizenship
for undocumented immigrants. May Day, the international worker’s holiday, has
become an occasion for immigrant communities in the United States to voice their
unique concerns (Johnston, 2012).
Some unions have worked to lessen discrimination against sexual minorities
(Cobble, 2007) and, since 1997, the AFL-CIO has recognized its LGBT mem-
bers as a formal constituency group. Pride at Work, the officially recognized
group, had 16 chapters by 2005. Thus, with regard to its efforts to combat dis-
crimination against sexual minorities, the U.S. labor movement “has shifted from
being part of the problem to being part of the solution” (Hunt & Boris, 2007,
p. 98).
There is a renewed focus on young and immigrant workers, low-wage work-
ers, part-timers, and temporary workers, who are often “invisible” (Hertz, 2010;
Simmons & Harding, 2009). Many of these are found in domestic service, not sub-
ject to most labor and safety laws, and in McJobs—minimum-wage and low-skill
service, hotel and retail jobs—that are generally regarded as “dead-end.” Such a
Psychology and Labor Unions 273
label tends to devalue the work, discourages efforts to improve conditions, and
is a source of feelings of shame. Average turnover in such jobs is often close to
100%, since these jobs are highly stressful and repetitive, with great uncertainty
in hours and schedules.
A study of young workers in supermarket unions found that the union had
succeeded in raising wages, increasing benefits, and providing protection from
mistreatment and from being fired (Tannock, 2004).
According to the International Labor Organization, the United States has the
largest share of low-paid jobs among all industrialized countries (cf. Reich, 2013).
These are jobs as fast-food cooks, hotel cleaners, cashiers, and hospital orderlies.
A full-time kitchen food prep worker earned less than $19,000 in 2012, cashiers
less than $21,000. The lowest paid workers are in limited-service restaurants
like McDonalds that employ more than 2 million workers, and workers in seven
of the largest corporations in the United States earn less than $30,000 a year.
Among personal home care aides, the fastest growing occupation, 57% live in
poverty (cf. Porter, 2012). Most minimum wage workers are adults (approximately
80%), disproportionately women with children, in large corporations, not small
companies (Eskow, 2013a).
Unions are continuing their historical efforts to “improve the lives of low-wage
workers and . . . reduce inequality” (Greenhouse, 2008, p. 242)—adding inequal-
ity stemming from age and lesser skills to that of ethnicity and gender. In collabora-
tion with social justice and civil rights groups, unions are sponsoring actions aimed
at raising wages and improving conditions for such workers. Illustrative of this
effort is a multi-restaurant strike by fast-food workers in New York City whose me-
dian pay is $9 an hour ($18,500 a year for a full-time worker) (Greenhouse, 2012c).
As noted by Bittman (2013, par. 5) “a rapidly increasing number of food industry
and other retail workers are now fighting for basic rights: halfway decent pay, a real
work schedule, the right to organize, health care, paid sick days, vacations and re-
spect.” The SEIU is providing advice and financial assistance in this extraordinary
effort. Another union-supported group, the National Domestic Workers Alliance,
has reported the results of a first-ever national survey of more than 2,000 nannies,
caregivers, and housecleaners in urban areas (Burnham & Theodore, 2012). Using
a participatory methodology, the investigators documented the abuse and exploita-
tion experienced by domestic workers, most of whom are women—immigrants and
people of color—who keep silent about exploitation because of fear of job loss and
deportation.
Since the early 2000s, university faculty unions have reached out to con-
tingent or adjunct faculty members for inclusion (e.g., see AAUP, 2012), in
order to provide job security, fair compensation, and access to benefits. My
university now has three chapters of the American Association of University
Professors (AAUP)—for full-time faculty, graduate assistants, and per course
instructors.
274 Lott
Proposals for Advancing Psychology’s Research Agenda
Related to our neglect of scholarship on social class are unrecognized implicit
negative assumptions about working class characteristics found in our literature
(see Lott, 2010, 2012; Lott & Bullock, 2007), as well as in the media and in political
arguments about persons who receive welfare and other so-called entitlements
(Bullock, 2013). Rose (2004) has written about judgments often made about
the intelligence of people who labor in service work or in the manufacturing of
products—work often described as “mindless” and considered to be “neck down”
rather than “neck up.” He notes that we tend to judge a person’s intelligence by
the work she or he does, and we assume that persons in certain occupations are
not as smart as we are.
Rose (2004, p. 197) argues that “most working men and women try to find
meaning in what they do—through the activity of the work itself or through
what their wages make possible outside of the workplace.” How accurate is this
statement for workers in diverse workplaces, and for “excluded workers”—those
who are outside the traditional protections offered to workers by practice or law
(Williams, 2012)? I present this as a proposal for new research. Anderson (2012)
alluded to the importance of pursuing such a question as we try to understand
“work-life fit.” He wrote, “American workers aren’t just looking for monetary
gain—they want balance and meaning in their work lives” (p. 9).
Another proposal for research can be credited to Richard Trumka (cf.
Richardson, 2011), president of the AFL-CIO. Most people in the United States,
he believes, are largely ignorant of labor union history. How do demographic
groups differ on such knowledge, and is knowledge of corporations greater,
lesser, or the same as knowledge about unions? What variables correlate with
information, beliefs and attitudes about collective bargaining? What are the
sources for information and beliefs about labor unions—in schools and in the
media?
We can examine other empirical questions: Are minorities of color and peo-
ple who have experienced negative economic events more pro-union than others
(Chang, 2003)? What psychological characteristics distinguish the 52% of Amer-
icans found in a recent Gallup poll to approve of unions (cf. Greenhouse, 2011)?
Dionne (2011) has written of “our cool indifference to the heroism of those who
go to work every day” (par. 13). How common is such indifference, and what are
its antecedents and correlates?
Our social psychological literature can help us to understand the phenomenon
of scapegoating—a highly relevant construct for an investigation of the current
targeting of public sector unions said to be responsible for the hard economic
times experienced by cities and states. We can also examine the factors/variables
(both individual and institutional) that encourage or discourage interest in union
membership—as illustrated by the work of Mellor (2009). His survey of nonunion
Psychology and Labor Unions 275
workers explored the influence of economic factors and self-evaluations. Ham-
mer and Avgar (2005) propose that consideration of both potential benefits and
potential costs (e.g., dues, adversarial relationship with employer) affect workers’
interest in joining unions.
There is considerable literature in psychology on the influence of employment
and a variety of job-related factors on family life, and attention has been paid to the
importance of work experiences for individuals and families. APA’s Public Interest
Directorate held a congressional briefing on “The Power of Work” that dealt with
the impact of joblessness and the positive effects of job satisfaction on relationships
and well-being (APA, 2011). Yet, we remain indifferent to research on the wide-
ranging contributions that labor unions make to such issues (Zickar, 2004). A
recent example is a paper on work–home interface (ten Brummelhurst & Baker,
2012) that discusses problems associated with working overtime and having to
face new tasks at work—with no mention of unions. There are exceptions, such as
a report of the impact of work on mental health that links anti-union developments
to increased workplace stressors, decreased autonomy, under-rewards, and distress
(Dobson & Schnall, 2009).
A number of areas in our science and application might profit considerably
from adding to the demographic information we typically obtain from study re-
spondents or clients, whether or not they are union members. This may prove to
be an important variable in investigations of such issues as work stress, retirement
plans, and areas of life satisfaction.
Should psychologists be concerned with the fact that nearly half of workers
in the United States get no paid sick days (Greenhouse, 2006)? Can we ignore
the consequences of this state of affairs for family well-being and ignore the data
documenting that children get well sooner if their parents can take time off from
work to care from them? Cities and states have begun to legislate on this issue,
with San Francisco; Washington, DC; Seattle; and Connecticut now guaranteeing
paid sick days for workers (Ness, 2012).
APA’s Office of Socioeconomic Status is a significant addition to our institu-
tional structure, and may be able to attract outside agency funding for an ambitious
research program focused on some of these questions. Such a program would serve
well to advance APA’s mission (APA, 2013a). We strive to obtain, and to apply,
“psychological knowledge to benefit society and improve people’s lives,” and
our goals include achievement of “social justice, diversity, and inclusion.” Surely,
then, we need to widen our vision and accept social class as a vital feature of
personal, family, and community identification and life. Doing so will encourage
recognition of the significance of such class-based institutions as organized labor,
and encourage us to study its efforts to benefit society and improve the quality
of personal lives—both directly and indirectly, through the advancement of civil
rights. We can open a dialogue that should broaden and enrich many areas of our
scholarship and application.
276 Lott
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B.J.Pol.S. 48, 1075–1091 Copyright © Cambridge University Press, 2016
doi:10.1017/S0007123416000302
First published online 5 December 2016
Labor Union Strength and the Equality of Political
Representation
PATRICK FLAVIN*
Amid growing evidence of ‘unequal democracy’ in the United States, labor unions can play a potentially
important role by ensuring that low-income citizens’ opinions receive more equal consideration when elected
officials make policy decisions. To investigate this possibility, this article evaluates the relationship between
labor union strength and representational equality across states and finds evidence that states with higher
levels of union membership weigh citizens’ opinions more equally in the policy-making process. In contrast,
there is no relationship between the volume of labor union contributions to political campaigns in a state and
the equality of its political representation. These findings suggest that labor unions promote greater
political equality primarily by mobilizing their working-class members to political action and, more broadly,
underscore the important role that organized labor continues to play in shaping the distribution of political
power across American society.
Keywords: political inequality; political representation; labor unions; public opinion; US states.
There is growing empirical evidence that the United States is an ‘unequal democracy’ in which
affluent citizens are more likely to have their preferences reflected in government policy decisions
compared to disadvantaged citizens.1 This representational inequality has arisen for a variety of
possible reasons: compared to citizens with low incomes, wealthier citizens are more likely to vote
in elections,2 contribute to political campaigns,3 have their opinions represented by organized
interests and professional lobbyists,4 and serve as elected officials in government.5 In response,
scholars and concerned citizens have begun to search for possible remedies for political inequality.
For most of the twentieth century, labor unions’ attempts to organize their workers for
political action were viewed as an important counterweight to the political power amassed
by business interests and the affluent.6 It is no surprise, then, that among the growing number
of studies that document the recent rise in both political and economic inequality in the
United States, nearly all point to the steady decline of labor union membership and influence as
a major cause.7 However, this conclusion is premised on the assumption that stronger labor
* Department of Political Science, Baylor University (email: patrick_j_flavin@baylor.edu). A previous
version of this article was presented at the 2016 meeting of the Southern Political Science Association in San
Juan, Puerto Rico. I thank Nicholas Carnes, Michael Hartney and Christopher Witko for helpful comments. Data
replication sets are available at https://dataverse.harvard.edu/dataverse/BJPolS, and online appendices are
available at http://dx.doi.org/doi:10.1017/S0007123416000302.
1 Bartels 2008; Flavin 2012; Gilens 2012; Gilens and Page 2014; Jacobs and Skocpol 2005; Rigby and
Wright 2013.
2 Verba, Schlozman, and Brady 1995.
3 Schlozman, Verba, and Brady 2012.
4 Schlozman and Tierney 1986.
5 Carnes 2013.
6 Goldfield 1987; Rosenfeld 2014; Smith 2000.
7 Bartels 2008; Gilens 2012; Gilens and Page 2014; Hacker and Pierson 2010; Kelly 2009; Schlozman,
Verba, and Brady 2012; Volscho and Kelly 2012. According to the US Bureau of Labor Statistics, in 2014 only
11.1 per cent of all wage and salary workers were members of a labor union, compared to 20.1 per cent in 1983.
https://doi.org/10.1017/S0007123416000302
mailto:patrick_j_flavin@baylor.edu
unions actually help promote greater political equality. Despite a deluge of conventional
wisdom that this assumption is true, to date there has been no empirical investigation of the
precise relationship between labor union strength and the equality of political representation.
To advance our understanding of the impact of organized labor in American politics, this article
uses the variation in labor union membership and campaign contributions across US states to examine
the relationship between labor union strength and the equality of political representation between rich
and poor citizens. Using public opinion measures from the National Annenberg Election Surveys
(NAES) and data on state policy outcomes, I uncover evidence that states with higher levels of union
membership weigh citizens’ opinions more equally in the policy-making process. In contrast, there is
no relationship between the volume of labor union contributions to political campaigns in a state and
the equality of political representation. These findings suggest that labor unions promote greater
political equality primarily by mobilizing their working-class members to political action, as opposed
to influencing elected officials directly through contributions to political campaigns.
BACKGROUND AND THEORETICAL EXPECTATIONS
Political scientists and political observers more generally have long warned that political
representation in the United States is tainted by an upper-class bias, such that wealthier citizens
have more influence over government policy decisions than the poor.8 But, as the American
Political Science Association Taskforce on Inequality and American Democracy lamented,
‘Unfortunately, political scientists have done surprisingly little to investigate the extent of actual
inequalities of government responsiveness to public opinion – that is, whether distinct segments
of the country exert more influence than others.’9 A series of recent studies has sought to correct
this problem and more fully understand unequal political influence in the United States. Most
notably, Bartels demonstrates that the opinions of affluent constituents strongly predict the
voting behavior of their senators, while the opinions of those with low incomes display little or
no relationship.10 In addition, Gilens collects data from thousands of individual public opinion
poll questions and finds that subsequent federal government policy decisions disproportionately
reflect the views of the affluent, and that this is especially true when the preferences of the rich
and poor diverge.11 Investigations of unequal political representation at the state level have
tended to come to similar conclusions.12 In short, there is a growing body of empirical evidence
that shows the stated opinions of citizens with low incomes tend to receive little attention in
government policy decisions compared to their more affluent counterparts.
Labor unions, as an organizing instrument for working-class citizens, have the potential to act
as a counterweight to the political power amassed by business interests and the affluent.13 This
potential arises, I argue, through two primary theoretical mechanisms. First, labor unions can
help offset the pronounced socio-economic bias in voter turnout and political activity more
generally14 if they are successful in mobilizing workers with lower socio-economic status to
political action.15 In one illustrative example of labor union mobilization efforts, the AFL-CIO
8 Dahl 1961; Schattschneider 1960.
9 Jacobs and Skocpol 2005, 124.
10 Bartels 2008. Ellis (2012) finds a similar income bias in political representation for members of the US
House of Representatives.
11 Gilens 2012.
12 Flavin 2012; Rigby and Wright 2013.
13 Goldfield 1987; Rosenfeld 2014; Smith 2000.
14 Leighley and Nagler 1992; Rosenstone and Hansen 1993; Verba, Schlozman, and Brady 1995.
15 Leighley and Nagler 2007.
1076 FLAVIN
(through its Committee on Political Education) routinely devotes considerable resources to
voter information, registration and turnout drives.16 Moreover, in many states, mandatory
collective bargaining laws provide public sector unions with subsidized benefits that can help
lower the costs of mobilizing their union members to political action.17 It is not surprising, then,
that previous research emphatically suggests that unions are successful at mobilizing their
members to higher levels of political activity and engagement.18 When combined with evidence
that voters’ opinions tend to be better represented by the policy decisions of elected officials
than non-voters’ opinions,19 and proof that the views of low-income citizens tend to be better
reflected in government policies when they turn out to vote at higher rates,20 jurisdictions with
higher levels of union membership (and, by extension, higher levels of union political
mobilization) are likely to display more egalitarian patterns of political representation.
Secondly, labor union organizations routinely insert themselves into political campaigns by
contributing money directly to candidates running for office21 who support union-friendly
policies (such as a higher minimum wage, more generous government health care and
retirement support, and a more progressive taxation system) that low-income and working-class
citizens, in general, also tend to support.22 Although political scientists have uncovered little
evidence that campaign contributions directly influence legislators’ roll-call votes, there is
ample evidence suggesting that contributions can exert sway behind the scenes by influencing
who legislators agree to meet with, what issues they focus on and how they allocate their scarce
time while in office.23 As a result of these efforts, there is likely to be a more equal weighting of
citizens’ political opinions in government policy decisions in jurisdictions where union
organizations are more actively involved in contributing to candidates for elected office who
support labor-endorsed policy positions.24
Despite the potentially important role that labor unions can play in promoting more equal
consideration of citizens’ political opinions in government policy decisions, to date this
question has received scarce empirical attention. Although related studies suggest that states
16 Sorauf 1988.
17 For example, public school teachers’ unions are almost always contractually entitled to receive a
comprehensive list of all district employees at the outset of each school year with detailed contact information
that can be utilized for political mobilization purposes (Flavin and Hartney 2015).
18 Ahlquist and Levi 2013; Asher 2001; Flavin and Radcliff 2011; Francia 2006; Leighley and Nagler 2007;
Radcliff 2001.
19 Griffin and Newman 2005.
20 Fellowes and Rowe 2004; Fowler 2013; Hill and Leighley 1992.
21 Ansolabehere, de Figueiredo, and Snyder 2003.
22 Franko, Tolbert, and Witko 2013; Gilens 2009.
23 Baumgartner et al. 2009; Hall and Wayman 1990; Langbein 1986; Makinson 2003; Powell 2012; Schram
1995; Witko 2006.
24 Importantly, a third way in which labor unions can potentially promote greater political equality is by
employing an ‘insider approach’ of lobbying elected officials and urging them to implement public policies that
are supported by organized labor (Facchini, Mayda, and Mishra 2011; Masters and Delaney 2005).
Unfortunately, there currently exists no comprehensive database of lobbying activity or expenditures by sector in
every state that would allow for empirical testing of this potential mechanism. However, the National Institute on
Money in State Politics is engaged in an ongoing effort to collect and catalog data on lobbying expenditures, and
currently reports preliminary data for sixteen states. When I construct a measure of labor union lobbying
spending as a share of total lobbying spending in a state and run a regression for those sixteen states using the
model specification in Table 2, the coefficient for labor union lobbying is not statistically different
from 0 (complete results are reported in Appendix Table A5). Across the sixteen states for which data are
available, the labor union lobbying expenditures measure correlates at 0.56 with the labor union membership
measure and at 0.16 with the labor union campaign contributions measure. Because data collection efforts in this
area are ongoing, this is a fruitful avenue for future investigation.
1077
with higher levels of union membership are more likely to implement liberal public policies,25
less likely to implement business-friendly policies,26 and tend to have lower levels of poverty
and income inequality,27 our understanding of the specific effect of labor unions on the
opinion–policy linkage between citizens and their government remains limited. Therefore, in
what follows, I take advantage of the variation in labor union membership and campaign
contributions across the American states to empirically evaluate the relationship between labor
union strength and the degree to which the political opinions of the wealthy and poor are
equally reflected in the policy decisions made by elected officials.
MEASURING THE EQUALITY OF POLITICAL REPRESENTATION IN US STATES
In this study, political representation is measured using a proximity technique that places public
opinion and policy on the same linear scale and compares the distance between the two.28 Using this
method, as the ideological distance between a citizen’s opinion and policy grows (that is, policy is
ideologically ‘further’ from a citizen’s preferences), that citizen is not well represented.29 In practical
terms, this proximity technique allows a researcher to evaluate whether a conservative (liberal)
citizen lives in a state that, compared to other states, implements conservative (liberal) policies and is
‘well’ represented, implements liberal (conservative) policies and is ‘poorly’ represented, or
gradations in between.
Measuring ideological proximity requires two pieces of data: (1) a measure of citizens’
opinions and (2) a measure of state policy. To measure public opinion, I combine data from the
2000, 2004 and 2008 NAES, three random-digit-dialing rolling cross-sectional surveys
conducted in the months leading up to that year’s presidential election. The major advantage of
pooling these three NAES surveys is their sheer sample size, which allows a large enough
sample without having to aggregate across a long time period or simulate state opinion.30 This
large sample size is especially important because this article later evaluates the relationship
between income and ideological proximity within individual states.31
Citizens’ general political ideology is measured using the following item from the NAES:
‘Generally speaking, would you describe your political views as very conservative, conservative,
moderate, liberal, or very liberal?’ The measure is coded from −2 (very conservative) to +2 (very
liberal). Data on citizens’ self-reported political ideology have been commonly used to measure
public opinion in previous studies of political representation,32 and there is reason to be confident
that self-reported ideology is an accurate measure of citizens’ aggregated policy-specific opinions.33
25 Radcliff and Saiz 1998.
26 Witko and Newmark 2005.
27 Brady, Baker, and Finnigan 2013; Kelly and Witko 2012.
28 Achen 1978; Ellis 2012; Flavin 2015.
29 The identical measurement technique has been used in several recent studies to evaluate the ideological
distance between citizens and members of Congress (Ellis 2012; Griffin and Flavin 2007; Griffin and Newman
2008), senators (Gershtenson and Plane 2007) and presidential candidates (Burden 2004; Jessee 2009) in the
United States, as well as the ideological distance between citizens and political parties in Europe (Blais and
Bodet 2006; Giger, Rosset, and Bernauer 2012; Golder and Stramski 2010; Powell 2009).
30 Carsey and Harden 2010.
31 A total of 177,043 NAES respondents across the three survey waves answered the ideological
self-placement and income items. All states except North Dakota (N = 475) and Wyoming (N = 414) have a
sample size of over 500 respondents. Alaska and Hawaii were not surveyed.
32 Bartels 2008; Erikson, Wright, and McIver 1993; Flavin 2012, 2015; Griffin and Flavin 2007.
33 For example, only 38 per cent of respondents who place themselves in the ‘very conservative’ category
believe that ‘Government should reduce income differences between the rich and poor’, while 77 per cent of
respondents who place themselves in the ‘very liberal’ category support that policy proposal. Similarly,
1078 FLAVIN
To measure public policy, I require a general measure of the ‘liberalism’34 of state policy
outputs that comports with the survey item that asks citizens their general political ideology. In
their seminal book on state opinion and policy, Erikson, Wright and McIver developed a
composite index of state policy liberalism using eight policy areas on which liberals and
conservatives typically disagree.35 Gray et al. updated this policy liberalism measure for 2000
using the following five policy items: (1) state regulation of firearms as measured by state gun
laws; (2) scorecard of state abortion laws in 2000; (3) an index of welfare stringency that
accounts for Temporary Assistance to Needy Families rules of eligibility and work requirements
for 1997–99; (4) a dummy measure of state right-to-work laws in 2001 and (5) a measure of tax
progressivity calculated as a ratio of the average tax burden of the highest 5 per cent of a state’s
earners to that of the lowest 40 per cent of a state’s earners.36 These five components are then
standardized and summed in an additive index such that more liberal state policies are coded
higher. I use this index as my first measure of the general ideological tone of state policy.
Secondly, a recent article by Sorens, Muedini and Ruger provides a rich source of data on
state policies in twenty different areas ranging from public assistance spending to gun control to
health insurance regulations.37 In addition to specific statutes and spending data, the authors
provide a summary index of policy liberalism for each state that they derive by factor analyzing
their entire range of policies. I use this composite score as a second measure of general policy
liberalism.38 Together, the two policy liberalism measures represent the unidimensional liberal/
conservative ideology of state policy decisions that correspond well to the measure of citizens’
general political ideologies described above.
Measuring ideological proximity requires a method of placing citizens’ opinions and state
policy on a common scale for comparison. Drawing on previous studies that have also used a
proximity technique to measure political representation,39 this article approaches this task in
three different ways. If all three measurement techniques point to the same conclusion, then we
can be more confident in the robustness of the results.40
First, all ideological opinions are standardized to a mean of 0 and a standard deviation of one,
and the two recent measures of general state policy liberalism described above41 are then
(F’note continued)
81 per cent of respondents who place themselves in the ‘very liberal’ category oppose ‘Laws making it more
difficult for a woman to get an abortion’, while only 28 per cent of respondents who place themselves in the ‘very
conservative’ category oppose that policy proposal. For additional examples, see Appendix Table A1.
34 Klingman and Lammers 1984.
35 Erikson, Wright, and McIver 1993.
36 Gray et al. 2004. They argue that using these policy items, as opposed to a measure of per capita
expenditures for different policy areas, precludes the possibility that policy liberalism is simply a proxy for a
state’s wealth. The five measures produce a Cronbach’s alpha of 0.63.
37 Sorens, Muedini, and Ruger 2008. The state policy data is available at http://www.statepolicyindex.com.
38 The Gray et al. (2004) and Sorens, Muedini, and Ruger (2008) policy liberalism measures correlate across
the states at 0.79.
39 Achen 1978; Blais and Bodet 2006; Burden 2004; Ellis 2012; Flavin 2015; Gershtenson and Plane 2007;
Giger, Rosset, and Bernauer 2012; Golder and Stramski 2010; Griffin and Flavin 2007; Griffin and Newman
2008; Jessee 2009.
40 One common critique of using the proximity method to evaluate political representation is that, regardless
of the statistical technique used to match up the two, opinion and policy are not on the same scale. However,
whatever the flaws of each of the three different measures of ideological proximity in matching up opinion and
policy, they are likely equally flawed for all citizens regardless of their income. Therefore, the proximity
measures are appropriate for evaluating how ideologically proximate opinion and policy are for a poor person in
comparison to a rich person (also see Ellis 2012).
41 Gray et al. 2004; Sorens, Muedini, and Ruger 2008.
Labor Union Strength and the Equality of Political Representation 1079
http://www.statepolicyindex.com
standardized as well. After standardizing both opinion and policy, they are now on a common
(standardized) metric, similar to the strategy originally used by Wright and more recently
by Ellis.42 Proximity is measured as the absolute value of the difference between a respondent’s
ideology score and the policy liberalism score for his/her state using both measures of policy.
This creates the first measure of ideological distance for each respondent in the NAES sample,
which is labeled the Standardized measure.
Secondly, the two measures of state policy are rescaled to the same scale (−2 to +2) as
citizens’ self-reported ideology. This technique is similar to that used in early studies of
congressional representation,43 and is still advocated by representation scholars today.44
The absolute value of the distance between a respondent’s ideology score and the policy
liberalism score for his/her state is again computed and labeled the Same Scale measure.
Thirdly, policy is rescaled to a tighter range (−1 to +1) than citizens’ ideologies, as suggested
and implemented by Powell in her studies of congressional representation.45 This procedure is
used because we can expect citizens’ ideological opinions to have a wider range and take on
more extreme values compared to actual state policy outputs. Again, the absolute value of the
distance between a respondent’s ideology score and the state policy liberalism score for his/her
state is computed and labeled the Restricted Scale measure.
Together, there are three different measurement techniques and two different measurements of
state policy liberalism, for a total of six different measures of ideological proximity between
citizens’ opinions and state policy. I am then interested in whether there are systematic differences in
proximity between opinion and policy across citizens; specifically, whether there is a link between a
citizen’s income and the ideological distance between opinion and policy. Because I am interested in
unequal political representation within each state and state populations can vary widely in terms of
their income distribution, it would be unwise to simply compare the incomes of citizens in one state
to those in another state. Simply put, we might expect someone making $100,000 per year living in
West Virginia to exert comparatively greater political influence than someone making $100,000 per
year living in Connecticut. To account for differences in income distribution across states, I generate
a measure of state relative income that compares a respondent’s income with the average income for
a resident in his or her state.46
With this measure of state relative income, I then assess whether there is a systematic
relationship between citizens’ incomes and the ideological distance between their opinion and
state policy by regressing the measure of ideological distance on income for every respondent in
the sample using the six different measures of ideological proximity described above.47
42 Ellis 2012; Wright 1978.
43 Achen 1978; Miller 1964.
44 Burden 2004; Giger, Rosset, and Bernauer 2012; Griffin and Newman 2008.
45 Powell 1982; Powell 1989.
46 The NAES codes respondents into one of nine categories based on their self-reported household income.
To create a state relative measure of income for each respondent, I take the respondent’s income category and
subtract the average income category value for all respondents in the state. The resulting calculation is positive if
a respondent’s income is above the state average and negative it is below the state average.
47 The results of these regressions are reported in Appendix Table A2. Because respondents are clustered
within states and experience the same state policy, I use standard errors clustered by state for all regressions. The
choice of a single time period (2000–08) for the analysis is primarily a function of data availability on public
opinion. As discussed above, while the NAES collects a large sample size in each wave, it is a national sample
that is not designed to sample equal numbers of respondents in each state. As a result, many small population
states have a sample size of less than 150 respondents in any single survey wave. Because the measure of the
equality of political representation developed below requires running a separate regression for each state, a
sample size of less than 150 survey respondents is problematic for the precision of the regression estimates.
1080 FLAVIN
The results of these six regression estimations reveal strong evidence of unequal political
representation. Specifically, all six coefficients for income are negative and bounded below 0,
which indicates that as a respondent’s income increases, the distance between their ideology and
state policy decreases and they are better represented. Put another away, the lower a
respondent’s income, the greater the distance between opinion and policy and the worse that
respondent’s general political ideology is represented in the general liberalism of his or her
state’s public policies.
Substantively, the larger opinion–policy distance for a respondent at the 10th percentile for
income compared to a respondent at the 90th percentile is about the same as the difference
between a respondent at the 10th percentile for (state relative) level of education compared to the
90th percentile48 and larger than the difference between an African-American respondent and a
white respondent.49 These findings comport with the small (but growing) set of studies that has
found that citizens with low incomes are systematically under-represented in the policy-making
process in US states.50
The primary rationale for examining unequal political representation at the state level is to
understand and explain variation in political equality across states. To assess in which states
political influence is strongly tied to income compared to those that weight opinions more
equally, I run a separate regression for each state and compare the coefficient for (state relative)
income. Similar to the nationwide regression discussed above, a more steeply negative slope
coefficient indicates a stronger relationship between income and ideological distance and,
accordingly, less political equality. For example, consider the two hypothetical states presented
in Figure 1. For each state, the line represents the slope of the relationship between income and
ideological distance. As the figure illustrates, the relationship between income and distance is
rather weak in State A, indicating that citizens’ opinions are weighted roughly equally
regardless of their income. In contrast, the slope of the relationship between income and
ideological distance is quite steeply negative for State B, indicating that there is a strong degree
of political inequality in state policy making.
A separate regression is run for each state using each of the six different measures of
ideological proximity described above (three measurement techniques × two measures of state
policy liberalism).51 When the six regression coefficients (for state relative income) are
(F’note continued)
By comparison, when pooling the 2000, 2004 and 2008 NAES waves together for the analysis, all states except
North Dakota (N = 475) and Wyoming (N = 414) have a sample size of over 500 respondents. Therefore, a
pooling strategy is necessary to have enough respondents to run state-specific regressions and evaluate the
relationship between income and opinion–policy proximity within each state. Nevertheless, as a robustness check
and to make sure pooling the data is not masking differences across survey waves, I also run a separate analysis
for each wave on the national sample and report the results in Appendix Table A3. As the regression coefficients
indicate, the relationship between income and ideological distance is negative (that is, state policy is more
proximate to citizens’ opinions as income increases) and of roughly the same magnitude for all three
survey waves.
48 Gilens 2005.
49 Griffin and Newman 2008.
50 Flavin 2012; Rigby and Wright 2013.
51 One potential concern with running a regression separately for each state with opinion–policy distance as the
dependent variable is that every respondent has the same value for state policy, effectively making the policy term a
constant. However, consider a state where income and ideological conservatism correlate perfectly (that is, as income
increases, so does ideological conservatism). If the state’s policy position is more conservative than all citizens’ ideology
positions, the regression coefficient for income would be negative (indicating that as income increases, the ideological
distance between opinion and policy decreases). But if the state’s policy position is more liberal than all citizens’
ideology positions, the coefficient for income would be positive (indicating that as income increases, the ideological
Labor Union Strength and the Equality of Political Representation 1081
compared across the states, they have a Cronbach’s alpha of 0.96, indicating that all six
measures appear to be measuring the same concept. To create a single summary score of
political equality that is directly comparable across states, I conduct a principal components
analysis on the six slope coefficients and generate a single factor score for each state.52 Because
a more steeply negative slope coefficient indicates more unequal representation (that is, a
stronger relationship between income and ideological distance), a more positive factor score
indicates greater political equality (that is, a more equal weighting of citizens’ opinions). I label
this new measure the Equality of Political Representation Index.
The factor scores generated using this procedure are reported in Table 1, where the states are
ranked from the most to least equal in terms of political representation. It is important to note
that the index is not simply an alternative measure of the general liberalism of state policy (with
the expectation that lower-income citizens support more liberal policies). The Equality of
Political Representation Index correlates with the Gray et al. policy liberalism measure at 0.46
and with the Sorens, Muedini and Ruger policy liberalism measure at only 0.36.53 Most
importantly, however, is the fact that there is significant variation in political equality across
states. In the following section, I use this variation to evaluate whether states with stronger labor
unions tend to display more egalitarian patterns of political representation.54
State A
State B
Respondent’s income
Ideological distance between a citizen’s opinion and state policy
Fig. 1. Computing the relationship for income and ideological distance, by state
Note: State A has more equal political representation than State B because the relationship (regression slope
coefficient) between income and opinion–policy distance is weaker in State A compared to State B.
(F’note continued)
distance between opinion and policy also increases). Even though the distribution of citizens’ opinions is identical under
both scenarios, the regression coefficients are very different depending on where state policy is located in the ideological
space (relative to citizens’ opinions). Therefore, the coefficient for respondents’ income for single-state regressions does
not simply indicate the relationship between income and ideology within a state but instead indicates (as intended) the
sign and strength of the relationship between income and opinion–policy distance.
52 The eigenvalue for the lone retained factor is 5.15 and explains 86 per cent of the total variance.
53 Gray et al. 2004; Sorens, Muedini, and Ruger 2008.
54 Although inequality in political influence among different racial/ethnic groups is not the focus of this article, it is
nonetheless possible that the Equality of Political Representation Index may correlate with state racial/ethnic
composition such that states with higher proportions of residents who are racial minorities are also those in which
citizens with low incomes are most unequally represented. To investigate this possibility, I ran a set of regression
models using the same specification as reported in Table 2 and added a measure of the percentage of state residents
that identifies as a race/ethnicity other than white (Caucasian). In all models, the coefficient for the percent of a state’s
1082 FLAVIN
STATE LABOR UNION STRENGTH AND THE EQUALITY OF POLITICAL REPRESENTATION
Above, I theorized that labor unions can promote greater political equality through two primary
theoretical mechanisms: (1) mobilizing union members to political action and (2) contributing to
candidates for elected office. To evaluate the individual effect of each mechanism, I measure the
strength of labor unions in a state in two different ways. First, to measure the potency of labor
mobilization in a state, I take the average percentage of nonagricultural wage and salary employees
(including employees in the public sector) who are union members for 2000–06 using data from
the Current Population Survey.55 Secondly, to measure labor union involvement in political
campaigns in a state, I take the amount of campaign contributions to candidates for state office
(governor, state senate and state house) for 2000–06 that come from labor unions56 and divide by
the total contributions from all sectors including agriculture, communications and electronics,
T A B L E 1 Ranking the States by the Equality of Political Representation
Montana 4.51 (most equal) Virginia 0.22
Minnesota 3.23 Florida 0.22
Oregon 3.19 Massachusetts 0.19
South Dakota 2.60 Connecticut 0.08
Vermont 2.19 Texas 0.01
California 2.18 Nevada −0.06
New Mexico 2.12 North Carolina −0.18
Michigan 1.94 Kansas −0.25
Washington 1.82 Maryland −0.50
Wisconsin 1.64 Kentucky −0.68
Ohio 1.54 New York −1.07
Nebraska 1.29 Indiana −1.27
Iowa 1.24 Louisiana −1.46
Pennsylvania 1.23 Tennessee −1.53
West Virginia 1.20 South Carolina −1.79
Arizona 1.15 Delaware −1.85
Missouri 1.14 North Dakota −2.02
Idaho 1.10 New Hampshire −2.36
Rhode Island 1.06 Arkansas −2.47
New Jersey 1.03 Oklahoma −2.52
Maine 0.57 Wyoming −2.91
Colorado 0.55 Georgia −3.56
Illinois 0.40 Alabama −5.06
Utah 0.34 Mississippi −8.44 (most unequal)
Note: cell entries are factor scores from combining six coefficients for state specific regressions.
Larger positive values indicate greater political equality (a weaker relationship between income and
ideological proximity).
(F’note continued)
residents that is non-white is not statistically different from 0, which indicates that more racially diverse states are no
more or less likely to exhibit unequal political representation along income lines. Moreover, the coefficients for the
other variables included in the regression models are substantive identical to the main findings reported in Table 2
below (complete results are reported in Appendix Table A6).
55 Hirsch, Macpherson, and Vroman 2001; Volscho and Kelly 2012. Union density data are available at http://
unionstats.gsu.edu/MonthlyLaborReviewArticle.htm. For 2000, private and public union membership rates
(density) correlated across the states at 0.73.
56 Included in this measure are contributions from general trade unions (construction, mining, etc.), public
sector unions (civil servants, teachers, etc.) and transportation unions (air, automotive, etc.).
Labor Union Strength and the Equality of Political Representation 1083
http://unionstats.gsu.edu/MonthlyLaborReviewArticle.htm
http://unionstats.gsu.edu/MonthlyLaborReviewArticle.htm
construction, defense, energy and natural resources, finance, insurance, real estate, general
business, health care, lawyers and lobbyists, and transportation.57 This calculation produces the
percentage of all campaign contributions (that are catalogued by industry) that come from labor
unions. Across states, these two measures of labor union strength correlate at 0.65.
In the analysis presented below, the Equality of Political Representation Index is regressed on
a state’s labor union membership and labor union campaign contributions to evaluate whether
states with stronger labor unions have more egalitarian patterns of political representation.58
Along with the two measures of labor union strength, I also include in the model a measure of
the partisan composition of state government, the composition of a state’s interest group
community, and a state’s median income and level of income inequality. The partisan
composition of state government is measured as the average percentage of Democrats in the
state legislature for 2000–06, and is included to account for the potential alternative explanation
that states with a higher percentage of Democratic legislators are both more likely to implement
union-friendly policies (that help strengthen labor unions) and more likely to equally represent
the political opinions of citizens with low incomes in state policy decisions. The composition of
a state’s interest group environment is measured as the percentage of organized groups in 1997
that represented for-profit interests (measure devised by Gray and Lowery and updated by
Gray et al.59) and is included because previous research indicates that a greater proportion of
for-profit interest groups attenuates the link between public opinion and state government policy
decisions.60 Finally, a state’s median income and level of income inequality (using the Gini
coefficient) for 2000 from the US Census Bureau are included because previous research on
unequal political influence at the state level suggests that political representation is the least
egalitarian in poorer states and in states with higher levels of income inequality.61
The analysis proceeds by including each of the two union strength variables in a regression
model separately before including both in the same model and allowing them to compete for
statistical influence. Column 1 of Table 2 reports the estimates with labor union membership as
the measure of union strength and reveals that the coefficient for membership is positive and
statistically different from 0. This finding indicates that states with a greater percentage of
workers who are labor union members tend to weigh citizens’ opinions more equally in the
policy-making process. By contrast, Column 2 reports the estimates with labor union campaign
contributions as the measure of union strength and reveals that the coefficient for contributions
is not statistically different from 0, indicating that there is no relationship between the
proportion of campaign contributions that comes from labor unions in a state and the equality of
political representation. Finally, Column 3 reports estimates with both measures of union
strength included in the model and, again, the coefficient for labor union membership is
57 State campaign contribution data by industry are collected by the National Institute on Money in State
Politics and are available at http://www.followthemoney.org/.
58 Because the income–ideological proximity slope coefficients are estimated rather than observed for the
states and have different levels of uncertainty (Achen 2005; Lewis and Linzer 2005), I also run Feasible
Generalized Least Squares regressions in the second stage using the six individual sets of state regression
coefficients from the first stage (instead of the combined Equality of Political Representation Index) as the
dependent variables and weight observations by the inverse of a coefficient’s standard error. These six
estimations yield substantially similar results to those reported in Table 2 (see Appendix Table A7). The results
are also similar when Huber–White robust standard errors are employed.
59 Gray and Lowery 1996; Gray et al. 2004.
60 Gray et al. 2004.
61 Rigby and Wright 2011, 2013. Nebraska has a non-partisan state legislature, and Alaska and Hawaii were
not surveyed in the NAES, so N = 47 for the regression estimations. Descriptive statistics for all variables
included in the analysis are reported in Appendix Table A4.
1084 FLAVIN
http://www.followthemoney.org/
statistically different from 0 while the coefficient for labor union campaign contributions is not.
In addition, the other covariates in the model reveal that (as previous studies have found) states
with higher levels of income inequality tend to be less politically equal. Somewhat surprisingly,
the coefficient for the percentage of Democrats in the state legislature is not statistically different
from 0, indicating that there is little evidence that the partisan composition of state government
is associated with the equality of a state’s political representation.62
From a substantive standpoint, the magnitude of the relationship between labor union
membership and the equality of political representation is quite large. Column 4 of Table 2
reports the standardized coefficients (the predicted change in terms of standard deviations of the
Equality of Political Representation Index when the independent variable in question is
increased one standard deviation) from the regression estimated in Column 3. As illustrated in
the table, labor union membership has the largest substantive relationship with the equality of
political representation of any predictor in the model. Specifically, a one-standard-deviation
increase in the percentage of a state’s workers who are union members corresponds to more
than half (0.53) a standard deviation increase in the Equality of Political Representation Index.
In summary, the data indicate that labor union membership is an important predictor of
representational equality in US states.
One potential concern about the findings reported in Table 2 is that the ability of labor unions
to mobilize their members to political action or influence policy making through contributions
T A B L E 2 Labor Union Strength and the Equality of Political Representation
(1) (2) (3) (4)
Labor Union Membership 0.182** – 0.234*** 0.53
[0.070] [0.084]
Labor Union Campaign Contributions – 0.032 −0.072 −0.19
[0.059] [0.066]
% Democrats in State Legislature −0.009 0.008 −0.010 −0.06
[0.026] [0.027] [0.026]
% Interest Groups For-Profit −0.076 −0.098 −0.090 −0.19
[0.063] [0.069] [0.064]
State Median Income −0.007 0.051 −0.004 −0.01
[0.057] [0.057] [0.057]
State Income Inequality −31.177* −29.037 −32.957* −0.29
[17.220] [18.557] [17.259]
Constant 18.098** 17.312* 20.062** –
[8.783] [9.594] [8.948]
R2 0.30 0.19 0.32 –
N 47 47 47 –
Note: dependent variable is the Equality of Political Representation Index (higher value indicates a
more equal weighting of citizens’ political opinions). Cell entries are ordinary least squares
regression coefficients with standard errors reported beneath in brackets. Column 4 reports the
standardized coefficients for the model in Column 3 (the predicted change in terms of standard
deviations of the Equality of Political Representation Index when the independent variable in
question is increased one standard deviation). *p < 0.10, **p < 0.05, ***p < 0.01 (two-tailed test)
62 This finding is consistent with Bartels (2008) and Rigby and Wright (2013), who demonstrate that both
Republican and Democratic candidates/elected officials are unresponsive to the political opinions of citizens with
low incomes.
Labor Union Strength and the Equality of Political Representation 1085
to political campaigns are not optimally operationalized. To investigate this concern, I estimate
a set of additional regression models with different measures of both union membership/
mobilization and union campaign contribution activity. To measure how successful labor unions
are at mobilizing their members to political action, I use validated voter turnout data from the
2008 Cooperative Congressional Election Study to calculate the share of the electorate in each
state that is from a union member household. To measure how active labor unions are in
contributing to political candidates compared to other organized interests, I calculate the ratio of
labor union contributions to business contributions in each state for 2000–06.63 Using these two
alternative measures of labor union strength and the same empirical strategy and model
specification as above, the results of the additional estimations are reported in Table 3. The
coefficients for these estimations are consistent with those reported in Table 2. Namely, labor
union mobilization is a significant predictor of greater representational equality in a state, while
labor union campaign contributions are not. Again, the analysis points to the conclusion that
labor unions’ ability to promote more egalitarian patterns of political representation lies in their
effectiveness in organizing and then mobilizing union members to political action as opposed to
contributing directly to state political campaigns.64
T A B L E 3 Robustness Check Using Alternative Measures of Labor Union Strength
(1) (2) (3) (4)
Union Household Share of Electorate 0.090** – 0.120** 0.44
[0.043] [0.049]
Labor Union/Business Ratio of Campaign Contributions – −0.190 −1.835 −0.20
[1.311] [1.404]
% Democrats in State Legislature −0.003 0.010 −0.006 −0.03
[0.026] [0.027] [0.026]
% Interest Groups For-Profit −0.062 −0.112 −0.071 −0.15
[0.067] [0.069] [0.067]
State Median Income 0.016 0.062 0.011 0.02
[0.056] [0.055] [0.056]
State Income Inequality −32.896* −29.837 −35.983** −0.32
[17.745] [18.641] [17.752]
Constant 16.676* 18.489* 19.007** –
[9.051] [9.700] [9.149]
R2 0.26 0.18 0.29 –
N 47 47 47 –
Note: dependent variable is the Equality of Political Representation Index (higher value indicates a
more equal weighting of citizens’ political opinions). Cell entries are ordinary least squares
regression coefficients with standard errors reported beneath in brackets. Column 4 reports the
standardized coefficients for the model in Column 3 (the predicted change in terms of standard
deviations of the Equality of Political Representation Index when the independent variable in
question is increased one standard deviation). *p < 0.10, **p < 0.05, ***p < 0.01 (two-tailed test)
63 Across the states, union membership and union household share of the electorate correlate at 0.87, and
union campaign contributions as a percent of the total and the union/business contributions ratio correlate at 0.85.
Data on campaign contributions from business are from the National Institute on Money in State Politics and
include general business advocacy associations (e.g., Chambers of Commerce) and individuals and groups
engaged in business services, manufacturing, gambling and casinos, food and beverage hospitality, lodging and
tourism, liquor and tobacco companies and sales, and retail sales.
64 Another potential concern with the findings presented in Table 2 is that equality of political representation is
driving labor union membership (i.e., more egalitarian political representation in a state leads to higher levels of
1086 FLAVIN
CONCLUSION
Political equality is a cornerstone of democracy. As Sidney Verba declares, ‘One of the bedrock
principles in a democracy is the equal consideration of the preferences and interests of all
citizens.’65 However, recent studies at both the national66 and state levels67 report that, across a
wide array of issue areas, affluent citizens are more likely to have their preferences reflected in
government policy decisions compared to disadvantaged citizens. In response to these
revelations, scholars and concerned citizens alike have begun to turn their attention to searching
for possible remedies for political inequality. For example, Flavin finds that patterns of political
representation are more egalitarian in states with stricter lobbying regulations and suggests
tighter restrictions on the registration and conduct of professional lobbyists as one viable avenue
of promoting more equal consideration of citizens’ political opinions.68 Carnes also shows that
citizens from working-class and low-income backgrounds are strikingly under-represented in
state legislatures across the nation and, in response, recommends specific programs to recruit
more blue-collar workers to run for elected office and advocate the opinions of disadvantaged
citizens within government.69
This article uses variation in labor union membership and campaign contributions across US
states to examine the relationship between labor union strength and the equality of political
representation between rich and poor citizens, and uncovers evidence that states with higher
levels of union membership (and where union households make up a greater share of the
electorate) weigh citizens’ opinions more equally in the policy-making process. In contrast,
there is no statistical relationship between the volume of labor union contributions to political
campaigns in a state and the equality of political representation. Taken together, these findings
suggest that labor unions promote greater political equality primarily by mobilizing their
working-class members to political action as opposed to influencing elected officials directly
through contributions to political campaigns.
As illustrated by recent high-profile battles over ‘right-to-work’ legislation in several states
that have traditionally supported labor-friendly policies such as Indiana, Michigan and
Wisconsin, the ability to organize and join a labor union has increasingly become a politically
contentious issue. Even in states without right-to-work laws, local governments (such as cities
and counties in Illinois and Kentucky) are attempting to bypass state labor policies by creating
right-to-work enclaves within their jurisdictions. The ultimate aim (and effect) of these efforts is
to lower union membership rates and substantially reduce the influence of labor unions in both
the workplace and the political arena. Given the empirical findings reported in this article, this
increasingly adversarial climate for organized labor will likely only exacerbate existing
(F’note continued)
union membership). To investigate this concern, I estimated a model with state union membership as the
dependent variable as a function of the Equality of Political Representation Index and three variables measured in
2000 that are common in the literature on the determinants of state union density: a dummy variable for whether
a state has a mandatory collective bargaining law for state public employees (as originally specified by Valletta
and Freeman 1988) and the percentage of private sector employees in a state that is in the construction and
manufacturing sectors (Hirsch, Macpherson, and Vroman 2001). The results of this estimation are reported in
Appendix Table A8; they reveal that the coefficient for the Equality of Political Representation Index is not
statistically different from 0.
65 Verba 2003, 663.
66 Bartels 2008; Gilens 2012.
67 Flavin 2012; Rigby and Wright 2013.
68 Flavin 2015.
69 Carnes 2013.
Labor Union Strength and the Equality of Political Representation 1087
inequalities in political influence between the rich and the poor. Therefore, citizens who are
concerned about rising levels of economic and political inequality should focus greater attention
on the important role that organized labor can still play in shaping the distribution of power in
American politics.
More broadly, this article suggests a fruitful avenue for future study of the relationship
between organized labor and the equality of political representation. Recent research examines
the correspondence between public opinion and party positions/government policies in
twenty-one parliamentary democracies and reports evidence that, similar to the United States,
affluent citizens are more likely than disadvantaged citizens to have their opinions
represented.70 Power Resources Theory posits that one of the most effective ways for
lower- and working-class citizens to exert greater political influence and secure public policies
favorable to them is by organizing and affiliating with a labor union.71 However, to date, there
has been no empirical evaluation of whether countries with stronger and more politically active
labor unions display more egalitarian opinion–policy linkages. Therefore future research should
take advantage of the sizable cross-national variation in union strength and activity to explicitly
examine whether the important role that labor unions play in promoting political equality in the
United States applies internationally as well.
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Labor Union Strength and the Equality of Political Representation 1091
© Cambridge University Press 2016
- Labor Union Strength and the Equality of Political Representation
BACKGROUND AND THEORETICAL EXPECTATIONS
MEASURING THE EQUALITY OF POLITICAL REPRESENTATION IN US STATES
Fig. 1Computing the relationship for income and ideological distance, by stateNote: State A has more equal political representation than State B because the relationship (regression slope coefficient) between income and opinion–policy distance is w
STATE LABOR UNION STRENGTH AND THE EQUALITY OF POLITICAL REPRESENTATION
CONCLUSION
1Bartels 2008; Flavin 2012; Gilens 2012; Gilens and Page 2014; Jacobs and Skocpol 2005; Rigby and Wright 2013.2Verba, Schlozman, and Brady 1995.3Schlozman, Verba, and Brady 2012.4Schlozman and Tierney 1986.5Carnes 2013.6Goldfield 1987; Rosenfeld 2014; Smi
References
Raymo, J.M., Warren, J.R., Sweeney, M.M., Hauser, R.M., & Ho, J-H. (2011). Precarious employment, bad jobs, labor unions, and early retirement. Journal of Gerontology: Social Sciences,
66B(2), 249–259, doi:10.1093/geronb/gbq106. Advance Access published on February 10, 2011.
© The Author 2011. Published by Oxford University Press on behalf of The Gerontological Society of America.
All rights reserved. For permissions, please e-mail: journals.permissions@oup.com.
249
Received March 22, 2010; Accepted December 21, 2010
Decision Editor: Merril Silverstein, PhD
CHANGES in the nature of employment in the United States since the mid-1970s are well documented, with
particular attention paid to increasing employment insecu-
rity, growth in jobs that do not provide pension or health
insurance benefits, and decline in unionized jobs (e.g.,
Kalleberg, 2000, 2009; Kalleberg, Reskin, & Hudson,
2000; Loveman & Tilly, 1988). A number of studies have
demonstrated that exposure to precarious employment and
“bad jobs” has negative implications for various outcomes,
including earnings, wealth, and health. This work has focused
primarily on young adulthood and midlife, despite compelling
theoretical reasons to believe that employment experiences
across the life course may have important implications for
understanding individual variation in experiences at older
ages (O’Rand, 1996a, 1996b; O’Rand & Henretta, 1999).
Theories of cumulative stratification provide insights into
the ways in which experiences and exposures across the life
course influence outcomes later in life. In short, this
perspective suggests that differentials in well-being and
employment circumstances within a cohort will increase
over time as initial levels of advantage (or disadvantage)
experienced by individuals interact with later opportunities
and experiences (Dannefer, 1987; O’Rand, 1996a, 1996b;
O’Rand & Henretta, 1999). In this way, employment experi-
ences such as involuntary job loss and lack of pension and
health care coverage may play an important role in shaping the
distribution of financial resources, health, and other dimen-
sions of well-being at older ages. The fact that these character-
istics are also well-established predictors of retirement timing
suggests that earlier employment experiences may be an
important indirect source of variation in age of retirement.
A better understanding of linkages between earlier em-
ployment experiences and retirement timing is particularly
important in the context of growing tension between the
long-term trend toward earlier retirement and recent efforts
to promote extended labor force participation at older ages.
Although the modal age of retirement in the United States is
now 62 years (Warner, Hayward, & Hardy, 2010), public
and private policy efforts have increasingly sought to
encourage and facilitate extended labor force attachment at
older ages in response to concerns about projected labor
force shortages, loss of skilled workers, difficulties in
financing pay-as-you go public pension and health care
systems, and ensuring the financial well-being of older
Americans (Burtless & Quinn, 2002; Morton, Foster, &
Sedlar, 2005; Munnell & Sass, 2008).
Theories of cumulative stratification suggest several,
potentially offsetting, ways in which precarious employment
and exposure to bad jobs across the life course may influ-
ence the likelihood of early retirement via temporally
proximate correlates of retirement. On the one hand, the
experience of job loss or exposure to bad jobs may reduce
Precarious Employment, Bad Jobs, Labor Unions,
and Early Retirement
James M. Raymo,1 John R. Warren,2 Megan M. Sweeney,3 Robert M. Hauser,1 and Jeong-Hwa Ho4
1Department of Sociology, University of Wisconsin–Madison.
2Department of Sociology, University of Minnesota, Minneapolis.
3Department of Sociology and California Center for Population Research, University of California, Los Angeles.
4Department of Sociology, National University of Singapore.
Objectives. We examined the extent to which involuntary job loss, exposure to “bad jobs,” and labor union member-
ship across the life course are associated with the risk of early retirement.
Methods. Using data from the Wisconsin Longitudinal Study, a large (N = 8,609) sample of men and women who
graduated from high school in 1957, we estimated discrete-time event history models for the transition to first retirement
through age 65. We estimated models separately for men and women.
Results. We found that experience of involuntary job loss and exposure to bad jobs are associated with a lower risk of
retiring before age 65, whereas labor union membership is associated with a higher likelihood of early retirement. These
relationships are stronger for men than for women and are mediated to some extent by pre-retirement differences in
pension eligibility, wealth, job characteristics, and health.
Discussion. Results provide some support for hypotheses derived from theories of cumulative stratification, suggesting
that earlier employment experiences should influence retirement outcomes indirectly through later-life characteristics.
However, midlife employment experiences remain associated with earlier retirement, net of more temporally proximate
correlates, highlighting the need for further theorization and empirical evaluation of the mechanisms through which
increasingly common employment experiences influence the age at which older Americans retire.
Key Words: Cumulative Stratification—Employment—Life Course—Retirement.
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RAYMO ET AL.250
the likelihood of retiring early via lower wealth accumula-
tion and limited pension benefits. On the other hand, these
same employment experiences might contribute to early re-
tirement via poor health and lower commitment to work in
late midlife. The fact that labor force changes since the mid-
1970s have disproportionately affected women (Kalleberg
et al., 2000) highlights the importance of examining whether
these posited indirect influences on retirement timing differ
by gender.
In this article, we use detailed employment history data
from the Wisconsin Longitudinal Study (WLS) to examine
how exposure to precarious employment, bad jobs, and
labor union membership across the life course are related to
the risk of early retirement (prior to age 65). We also assess
the extent to which these relationships are mediated by
established correlates of retirement timing (such as private
pension eligibility, wealth, and health) and examine the
extent to which linkages between employment history and
early retirement differ by gender.
Background
The Changing Nature of Employment
A key feature of changes in the nature and quality of
employment since the mid-1970s is declining job security
(Hacker, 2008). Kalleberg (2009, p. 5) argues that layoffs
have become a basic component of employers’ restructur-
ing strategies over the past thirty years and that precarious
employment has spread from unskilled, less educated seg-
ments of the labor force to all sectors of the economy. Invol-
untary job loss is a widely-studied indicator of employment
insecurity, and recent research shows that the proportion of
employees who experience involuntary job loss in a given
three-year period has fluctuated between 9% and 13% since
the early 1980s (Farber, 2005). Estimates from the Bureau
of Labor Statistics indicate that more than 30 million invol-
untary job losses occurred between the early 1980s and
2004 (Kalleberg, 2009, p. 2).
Closely related to the increase in precarious employment
is growth in bad jobs. The quality of jobs can be evaluated
in a variety of ways, but bad jobs are frequently defined as
those that do not offer private pension plans or health insur-
ance benefits and are characterized by low wages (Hacker,
2008; Kalleberg et al., 2000; Mishel, Bernstein, & Allegretto,
2007). Recent studies show that 21% of employed Ameri-
cans did not have private health insurance coverage in 2008
(Turner, Boudreaux, & Lynch, 2009) and that half did not
have private pension coverage in 2004 (Munnell & Perun,
2006). In their tabulations of data from the 1995 Current
Population Survey, Kalleberg and colleagues (2000) show
that 60% of American workers had a job that was bad on at
least one indicator (no pension, no health insurance, or low
wages) and that one in seven had a job that was bad on all
three indicators.
A third widely-studied component of changes in the
nature of employment is decline in unionized jobs. The pro-
portion of employees in labor unions fell from .28 in 1970
to .12 in 2007 (Statistical Abstract of the United States,
1980 and 2009 volumes). Because unionized jobs are more
likely to provide employment security, benefits, and higher
pay, declining union membership is closely related to the
increase in precarious employment and bad jobs (Kalleberg
et al., 2000).
Explanations of these changes have emphasized increas-
ing global competition, technological change, deregulation,
the ability to outsource low-skill work to low-wage countries,
and the shift from manufacturing to service and information
industries, all of which have prompted employers to seek
more flexible employment relations (Hacker, 2008; Hipple &
Stewart, 1996; Tilly, 1996). This flexibility has been achieved
by moving from long-term employment to an increased reli-
ance on nonstandard work arrangements, including tempo-
rary/fixed-term employment and part-time employment,
which allow for reductions in the cost of providing benefits
and facilitate the expansion or contraction of employee
numbers in response to shifting demand (Farber & Levy,
2000; Hipple & Stewart, 1996; Kalleberg, 2009; Tilly,
1996).
Relationships Between Midlife Employment Experiences
and Early Retirement
Theories of cumulative stratification suggest several
reasons to expect that the quality of employment across
the life course has important implications for when older
Americans retire. Of particular importance is evidence that
precarious employment, job loss, and employment in bad
jobs are negatively associated with various dimensions of
well-being, including earnings (Farber, 2005; Stevens, 1997),
wealth (Chan & Stevens, 2001), physical health (Strully,
2009), and psychological health (Burgard, Brand, & House,
2007; Grzywacz & Dooley, 2003). Involuntary job loss has
also been linked to reductions in subsequent labor force
participation and a higher likelihood of subsequent employ-
ment in part-time jobs (Farber, 2005). Job loss in late
midlife, in particular, has been linked to earlier retirement
(Chan & Stevens, 2004).
Because financial well-being, health, and employment
circumstances are also well-established correlates of retire-
ment age, processes of cumulative stratification provide an
important theoretical link between earlier employment
experiences and the timing of retirement. Developing an
empirical understanding of this link is particularly important
in light of growing efforts to reverse the long-term trend
toward earlier retirement and prolong labor force attachment
at older ages. Efforts to project the retirement behavior of
the baby boom cohort, to respond to concerns about the loss
of skilled older workers and pressures on social security,
and to ensure the financial well-being of older Americans
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PRECARIOUS EMPLOYMENT, BAD JOBS, LABOR UNIONS, AND EARLY RETIREMENT 251
will benefit from an understanding of the extent to which
exposure to increasingly common employment experiences
across the life course is associated with the likelihood of
early retirement.
Hypothesized relationships are mixed. On the one hand,
earlier exposure to precarious employment and bad jobs
may reduce the likelihood of early retirement by limiting
the financial capacity to retire. We posit that involuntary job
loss and employment in bad jobs will contribute to a lower
likelihood of early retirement via more limited access to
private pension coverage and lower wealth accumulation
(Crystal & Shea, 1990; O’Rand & Shuey, 2007). We also
expect the risk of early retirement to be lower among those
who did not belong to a labor union and are thus less likely
to rehired following layoff, to be eligible for employer-
provided pension benefits, or to face the institutionalized
incentives for early retirement present in many unionized
jobs (Hardy, Hazelrigg, & Quadagno, 1996). This scenario
is consistent with recent evidence that the long-term trend
toward earlier retirement has slowed or reversed (Friedberg,
2007) and that the economic feasibility of early retirement
is declining. Indeed, recent projections suggest that about
one-third of the early baby boom cohort (born 1946–1954)
will not be able to maintain their pre-retirement standard of
living even if they work full-time until age 65 (Munnell,
Golub-Sass, Perun, & Webb, 2007), and attitudinal survey
data indicate that increasing proportions of Americans
approaching retirement cite economic necessity as a reason
for their plans to continue working beyond current peak
ages of retirement (AARP, 2004; Merrill Lynch, 2006).
On the other hand, exposure to precarious employment
and bad jobs is expected to promote early retirement via
poor health and more marginal labor force attachment. Ear-
lier studies have found that unstable careers and spells of
unemployment across the life course are associated with
earlier retirement for men via higher rates of disability as
well as weaker attachment to work and a lower likelihood of
engaging in rewarding work (Hayward, Friedman, & Chen,
1998). Exposure to involuntary job loss and bad jobs across
the life course may also be associated with a higher likeli-
hood of experiencing unforeseen events such as health
decline or late-career job loss that result in relatively early,
and perhaps involuntary, retirement (Chan & Stevens, 2004).
The rise in precarious employment and the spread of bad
jobs has been particularly pronounced among women
(Kalleberg et al., 2000), suggesting that there may be im-
portant gender differences in relationships between earlier
work experiences and retirement timing. Again, theoretical
expectations are ambiguous. The effect of earlier employ-
ment experiences may be particularly strong for women
who are more likely to experience career interruptions dur-
ing prime childbearing years that can result in shorter em-
ployment tenure and a higher likelihood of employment in
part-time nonstandard jobs that do not provide pension and
health insurance benefits (O’Rand, 1988). Alternatively, it
is possible that earlier employment experiences are less
relevant for women whose retirement decisions may be more
sensitive than men’s to family circumstances such as care-
giving obligations and to the characteristics and circum-
stances of their spouses (Szinovacz & DeViney, 2000).
Our objective in this article was to extend our under-
standing of linkages between work experiences across the
life course and early retirement using data from the WLS.
In particular, we examine how exposure to precarious em-
ployment, bad jobs, and labor union membership across the
life course are related to the risk of early retirement (prior to
age 65) and assess the extent to which these relationships
are mediated by established correlates of retirement timing.
In doing so, we make three important empirical and theo-
retical contributions. First, the WLS contains rich life his-
tory data for one of the first cohorts to approach retirement
after facing increased exposure to precarious employment
and bad jobs for a large proportion of their working lives,
thus making it a uniquely valuable source of insight into the
long-term implications of the changing nature of employment
over the past three decades. Second, the WLS data allow for
improvement on earlier studies on linkages between work
experiences and variation in retirement outcomes (e.g.,
Hayward, 1986; Hayward, Grady, Hardy, & Sommers,
1989; Hayward et al., 1998; O’Rand & Henretta, 1982;
O’Rand & Landerman, 1984), which are limited by their
reliance on older surveys such as the National Longitudinal
Surveys of the late 1960s, their focus on men, and their con-
sideration of relatively limited aspects of work histories
(e.g., characteristics of longest job). Finally, we extend
recent life course theorization by evaluating hypotheses
about the ways in which processes of cumulative stratifica-
tion influence early retirement, one important aspect of the
increasingly heterogeneous retirement process (e.g., O’Rand,
1996a, 1996b; O’Rand & Henretta, 1999).
Data and Methods
Sample
The WLS is a long-term study of a random sample of
10,317 men and women who were first interviewed prior to
graduating from Wisconsin high schools in 1957 and rein-
terviewed in 1975, 1993, and 2004. Several features of the
WLS make it ideal for our purposes. First, it provides de-
tailed information on almost all jobs that respondents have
held between ages 35 and 65, allowing us to measure work
histories in far greater detail than in previous studies. Sec-
ond, the WLS respondents were 35–36 years old in 1975,
making this survey a rich source of information on the em-
ployment histories of one of the first cohorts exposed to the
rise in precarious employment and bad jobs and the decline
in union membership from early career through peak
retirement ages. Third, the WLS cohort is one of the first
in which significant numbers of women have worked
throughout the life course, allowing us to examine gender
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RAYMO ET AL.252
differences in relationships between earlier work experi-
ences and retirement.
Our analyses are based on data collected in the 1993 sur-
vey (conducted in 1992–1994 with a response rate of 87%)
and the 2004 survey (conducted in 2003–2005 with a
response rate of 85%). We excluded 1,593 members (15%)
of the original cohort who did not participate in either the
1993 phone interview (when they were 53–54 years old) or
the 2004 phone interview (when they were 64–65 years old)
and six respondents who either had missing information on
age or were several years older than the rest of the Class of
1957 graduates. These restrictions leave us with a base sam-
ple of 8,718 individuals. Auxiliary analyses indicate that
those with only a high school degree were more likely to be
among the 1,593 who were excluded due to refusal to par-
ticipate in either survey (n = 1,048) or death prior to the
1993 survey (n = 545). Our results may thus be biased to the
extent that relationships between unobserved earlier work
experiences and retirement outcomes are systematically
different for these missing cases.
It is also clear that the likelihood of being censored in
1993 due either to refusal to participate in the 2004 survey
(n = 879) or to death between 1993 and 2004 (n = 578) was
somewhat higher among those with less education and lon-
ger exposure to bad jobs. Because we have no information
about these individuals’ experiences after the 1993 inter-
view (other than year and month of death for those who
died), we censored these observations in the year of the
1993 survey. To assess the potential impact of this attrition,
we estimated auxiliary models based on different assump-
tions about the unobserved experiences of this group subse-
quent to the 1993 survey. In particular, we assumed that (a)
all retired in the year following the 1993 survey, (b) all
retired in 2004 (for those lost to follow-up) or in the year of
death (for those who died), (c) all retired halfway between
the 1993 survey year and either the year of the 2004 survey
or the year of death, and (d) none retired prior to 2004 or
year of death. In all cases, we assume that values of inde-
pendent variables remained constant from 1993 until the
year of retirement or censoring. In all but one scenario, esti-
mated coefficients for indicators of precarious employment,
exposure to bad jobs, and belonging to a labor union were
robust to these alternative treatments of data missing due to
attrition following the 1993 survey. We note the one excep-
tion later when discussing the results.
Variables
Retirement timing.—In both the 1993 and 2004 surveys,
respondents were asked whether they considered them-
selves to be partly retired, completely retired, or not retired
at all. Those who were partly or completely retired were
asked the month and year in which they retired. For 178
respondents who did not report retirement status in either
survey, or reported being retired but did not provide a date
of retirement, we used other sources of information to de-
termine retirement timing. Reports of retirement as a reason
for leaving specific employment spells and information on
age of initial receipt of private pension benefits and social
security benefits enabled us to construct measures of retire-
ment experience and timing for 159 of those with missing
data on self-stated retirement. We excluded the remaining
19 (0.02% of the total sample) who provided no usable in-
formation on retirement. We also excluded 90 respondents
whose reported retirement age was younger than their age
in 1975, most (89%) of whom were women with very young
reported ages of retirement (the mean age of retirement for
this group was 26). Our final analytical sample is thus 8,609
(i.e., 8,718 − 19 − 90 = 8,609) or 88% of the 9,730 sample
members who survived to the 1993 survey.
We began by constructing a data file consisting of person-
year records in which each respondent contributed one
observation per year from 1975 through the year of first
self-stated retirement or the 2004 survey, whichever came
first. As noted previously, those who did not participate in
the 2004 survey were censored in the year of the 1993 sur-
vey. By constructing these person-year data, we were able
to easily incorporate time-varying information on pension
eligibility, job characteristics, family circumstances, and
health status collected in the 1993 and 2004 surveys and
to flexibly specify the baseline hazard of retirement in
discrete-time event history models.
Midlife employment experiences.—We constructed five
measures of midlife work experience that reflect cumulative
exposure to precarious employment, bad jobs, and union-
ized jobs. Except for the measure of labor union member-
ship, these variables were constructed from the employment
history data collected at ages 53–54 in the 1993 survey and
at ages 64–65 in the 2004 survey. Employment histories in
the WLS surveys are comprised of multiple employment
spells—uninterrupted periods of time working for the same
employer, including self-employment. These data provide
detailed information on employment status and the charac-
teristics of jobs held by respondents during each year
between 1975 and 2004. This information includes the
years that respondents started and stopped working for that
employer, the reason for ending that employment spell,
whether they worked full- or part-time, the industry and oc-
cupation when they began working for that employer, health
insurance coverage, and pension coverage.
Our measure of exposure to precarious employment is a
time-varying indicator of whether respondents ever left a
job involuntarily prior to first retirement. Based on open-
ended responses to questions asking the reason for ending
each employment spell, we define involuntary job loss as
cases in which the reported reason was business closing,
downsizing, relocation, termination, or layoff. This mea-
sure is equal to zero in all person-years prior to the year in
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PRECARIOUS EMPLOYMENT, BAD JOBS, LABOR UNIONS, AND EARLY RETIREMENT 253
which involuntary job exit was reported (if any) and equal
to one in that year and all subsequent years. For the small
number of respondents who report multiple job losses, we
focus on the most recent experience. Based on evidence that
job loss in late midlife is a particularly import predictor of
early retirement (Chan & Stevens, 2004), we define two
categories of involuntary job loss to distinguish those that
occurred at younger ages (before age 50) from those that
occurred in late midlife (age 50 and older).
To operationalize exposure to bad jobs, we constructed
time-varying indicators of the cumulative proportion of
working years since 1975 spent in jobs that (a) did not offer
private pension plans, (b) did not offer health insurance
coverage, and (c) were characterized by low earnings. As
discussed previously, the primary objective of this study
was to advance our understanding of the implications of
structural changes in the labor market by examining rela-
tionships between exposure to bad jobs across the life
course and early retirement. We therefore use information
on whether each reported job offered private pension and
health care coverage (rather than whether respondents par-
ticipated in such programs) and follow the measurement
strategy employed by Kalleberg and colleagues (2000) in
using information on job-specific average wages (rather
than the wages of individual respondents). We define
low-wage jobs as those that fall below the median value of
occupational earnings—the percentage of people in a given
occupation who reported hourly wages of at least $14.30 in
the 1990 census (Hauser & Warren, 1997). For each of these
three indicators of bad jobs, we calculated time-varying
measures of exposure by dividing the cumulative number of
years employed in a bad job (since age 35) by the cumula-
tive number of years employed since age 35. At any given
age, these measures thus indicate the cumulative proportion
of working years since age 35 spent in bad jobs.
Due to the limited information on labor union member-
ship available in the surveys, our measure of belonging to a
union is not truly a time-varying variable. It is constant with
a value of one for those who reported that they belonged to
a labor union in 1975, constant with a value of zero for
those who did not report union membership in either 1975
or 1993, and changes from a value of zero to a value of one
in 1993 for those who reported union membership in 1993
but not in 1975. Because these questions referred only to the
jobs held in 1975 and at the time of the 1993 survey, we
cannot identify labor union membership for all jobs held
since age 35.
Correlates of retirement timing.—Drawing on theories of
cumulative stratification, we attempt to evaluate pathways
through which earlier exposure to precarious employment,
bad jobs, and unionized jobs may be associated with retire-
ment outcomes by including time-varying contemporane-
ous measures of financial well-being, job characteristics,
employment status, and health. We lag these measures by
one year so that respondents’ economic, employment, and
health statuses at age t−1 are used to predict the likeli-
hood of retirement at age t. In contrast, the measures of
unexpected job loss, exposure to bad jobs, and union
membership (which are also lagged by one year) summa-
rize cumulative employment experiences between age 35
and age t−1.
We measured financial well-being using information on
private pension eligibility and wealth. Pension eligibility is
a time-varying trichotomous indicator that distinguishes
those who are eligible for private pension benefits at a given
age from those who are covered but not yet eligible and
those who are not covered by an employer-sponsored pen-
sion plan. Wealth is a time-constant indicator of respon-
dents’ percentile rank in the within-sample distribution of
net worth at the time of the 1993 survey. Net worth in the
WLS is constructed as the reported value of home equity,
real estate, business or farm, motor vehicles, savings, and
investments owned by the respondent and his/her spouse
minus their reported debt. Because available information on
net worth refers only to the time of the 1993 and 2004 sur-
veys, it is impossible to construct time-varying measures
without making untestable assumptions about how wealth
changes over time. We therefore make the simplest assump-
tion of no change over time in respondents’ relative position
in the within-sample distribution of net worth.
Job characteristics include time-varying indicators of
occupational sector of the current or last job (private sector,
public sector, self-employed), employment status (full-time,
part-time, not employed), occupational socioeconomic
standing, and duration of the current (non-) employment
spell. Occupational socioeconomic standing is measured as
the value of occupational education—the percent of persons
in each occupation in the 1990 Census who completed one
year of college or more (see Hauser & Warren, 1997, for
details).
Health status is measured using time-varying indicators
of four serious illnesses or health events after age 35. These
measures are constructed from information in the 2004 sur-
vey on the diagnosis of diabetes, cancer, heart problems,
and stroke and are equal to zero in person-years prior to the
experience of these health problems and equal to one in the
year of diagnosis and all subsequent years. Because
the 1993 survey did not ascertain the year of diagnosis, we
assume that these health events occurred midway between
1975 and the 1993 survey for the small number of respon-
dents who reported a diagnosis in 1993 but did not
respond to the 2004 survey. Among the 1,457 respondents
censored at the time of the 1993 survey, 75 reported a diag-
nosis of diabetes, 77 reported heart trouble, and 33 reported
cancer. The 1993 survey did not ask about experience of
stroke so we set this variable equal to zero for those who did
not respond to the 2004 survey. We also set all four health
indicators equal to zero for 558 respondents who did not
respond to the part of the 1993 survey that was conducted
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RAYMO ET AL.254
by mail (which contained the questions on these health
events).
Models
We examine relationships between earlier work experi-
ences and retirement through age 65 by estimating discrete-
time event history models using logistic regression. These
models can be expressed generally as:
α β γ ε− −− = + + +1 1 itln[ /1 ]it it it itp p X Y (1)
α β γ δ ε− − −− = + + + +1 1 1ln[ / 1 ] ,it it it it it itp p X Y Z (2)
where pit is the probability that the ith respondent retires
at age t, conditional on being not retired at age t−1. Xit−1
includes age, a time-invariant measure of educational at-
tainment (high school, some college, bachelor’s degree, and
graduate or professional degree), and a time-varying indica-
tor of marital status (married, divorced/separated, widowed,
and never married). Yit−1 includes the five employment
experience measures of central interest. The subscript t−1
indicates that independent variables are lagged one year. Two
exceptions are age and pension eligibility, both of which are
measured at age t. By including the correlates of retirement
timing described previously (Zit−1 in Model 2), we can eval-
uate the extent to which relationships between earlier
employment experiences and retirement timing change
after controlling for factors posited to delay retirement
(lower wealth, lack of pension eligibility, shorter employ-
ment tenure, and self-employment) and factors posited to
promote early retirement (lower occupational status, limited
labor force attachment, and health problems). We estimate
the two models separately for men and women in light of
the posited gender differences described previously.
Based on the results of preliminary descriptive analyses,
we specify the baseline hazard of retirement using a linear
term in combination with age-specific dummy variables
between ages 60 and 65. As in other studies using different
data and different methods, we find that the risk of retire-
ment rises through age 60, increases rapidly to a peak at age
62, and declines subsequently. We do not see a second peak
at age 65 because age in a given person-year refers to re-
spondents’ birthday in that calendar year and many of those
coded as age 65 had not yet reached their 65th birthday by
the survey date.
Results
Table 1 describes the analytic sample by sex in two ways.
Columns 1 and 2 describe the characteristics of individuals
(observed in the year prior to retirement or censoring) and
columns 3 and 4 present summary statistics for all person-
year observations. Individual data show that 64% of both
men and women had retired by the time of the 2004 survey
(when they were 63–65 years old). Among those who
retired, 95% retired before age 65 (not shown). The mea-
sures of midlife work experiences indicate that the propor-
tion of working years that women spent in jobs without
private pension coverage (.43), without health insurance
coverage (.35), and with low earnings (.64) is substantially
higher than that for men. The proportion with experience of
involuntary job loss is similar for men (.20) and women
(.17), but men were somewhat more likely to report an
involuntary job loss in late midlife. Women were less likely
to report belonging to a labor union (.18 vs. .33 for men).
Other measures indicate that men are more highly educated;
less likely to be divorced or widowed; more likely to be
eligible for private pension benefits; more likely to be self-
employed; much less likely to be employed part-time or not
working; and have higher net worth, longer employment
tenure, and somewhat higher occupational socioeconomic
standing. The prevalence of major health problems is low
and gender differences are small.
Table 2 presents exponentiated coefficients (odds ratios)
from two discrete-time event history models for the transi-
tion to retirement through age 65. Looking at the results for
Model 1, it is clear that our measures of exposure to increas-
ingly common employment experiences are associated with
a significantly lower risk of retirement (or equivalently, a
lower probability of retiring by age 65). This is especially
true for men, with four of the five employment history
measures negatively associated with the risk of retirement.
Specifically, the odds of retirement at a given age are about
one-third lower for men with earlier experience of involun-
tary job loss (regardless of when it occurred), 51% lower for
those who were consistently employed in jobs that did not
provide private pension plans (relative to those who always
had access to pension coverage), 17% lower for those who
were always employed in jobs characterized by low earn-
ings, and 47% higher for those who belonged to a labor
union (or, equivalently, 32% lower for those who did not
belong to a labor union). The proportion of working years
spent in jobs that do not provide health insurance coverage
is not significantly related to the risk of retirement.
In Model 2, measures of net worth, pension eligibility,
occupational sector, spell duration, occupational socioeco-
nomic standing, employment status, and health problems
are all significantly related to men’s risk of retirement in
expected ways. More importantly, the inclusion of these
temporally proximate correlates of retirement attenuates,
but does not eliminate, the statistically significant relation-
ships between earlier employment experiences and the risk
of retirement. The only change is that the coefficient for in-
voluntary job loss prior to age 50 is no longer statistically
significant at p < .05 (p = .052). Furthermore, the positive
coefficient for longer exposure to jobs without health insur-
ance benefits now approaches statistical significance (p =
.09), a relationship that emerged after controlling for
self-employment, a strong predictor of both limited access
to health care across the life course and later retirement.
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PRECARIOUS EMPLOYMENT, BAD JOBS, LABOR UNIONS, AND EARLY RETIREMENT 255
One plausible explanation for this finding is that those with
limited health insurance coverage across the life course may
need to take another (postretirement) job that provides
health insurance until becoming eligible for Medicare.
However, supplementary analyses reveal that this is not the
case. In most cases, early retirement for this group is fol-
lowed by self-employment.
In only one case were the results presented in Table 2 sen-
sitive to the inclusion of imputed observations for men who
did not respond to the 2004 survey. When we assumed that
none of these men retired and were censored at the year of the
2004 survey or the year of their death, coefficients for invol-
untary job exit and not belonging to a union were no longer
significantly different from zero. When we make less extreme
assumptions about the unobserved behavior of this group, the
results are very similar to those presented in Table 2. We are
thus inclined to believe that our results are reasonably robust
to nonrandom loss to follow-up and death.
For women, like men, results of Model 1 indicate that
experience of involuntary job loss (after age 50) is associated
Table 1. Sample Characteristics, by Sex
Variable
Individualsa Person-years
Men Women
Men Women
Retired
No 0.36 0.36 0.97 0.97
Yes 0.64 0.64 0.03 0.03
Age 58.36 (5.17) 57.88 (5.69) 48.68 (7.54) 48.51 (7.58)
Educational attainment
High school 0.55 0.69 0.54 0.69
Some college 0.14 0.13 0.15 0.13
Bachelor’s degree 0.14 0.12 0.14 0.12
Graduate/professional degree 0.17 0.06 0.17 0.06
Marital status
Married 0.85 0.77 0.87 0.82
Divorced/separated 0.09 0.13 0.08 0.11
Widowed 0.01 0.06 0.01 0.03
Never married 0.05 0.04 0.05 0.05
Midlife work experiences
Involuntary job exit
None 0.80 0.83 0.89 0.90
Before age 50 0.08 0.08 0.06 0.05
After age 50 0.12 0.09 0.06 0.04
Proportion of years w/o access to private pension 0.26 (0.37) 0.43 (0.42) 0.27 (0.40) 0.42 (0.45)
Proportion of years w/o access to health insurance 0.15 (0.30) 0.35 (0.41) 0.14 (0.31) 0.34 (0.43)
Proportion of years in low earning occupation 0.34 (0.42) 0.64 (0.43) 0.34 (0.44) 0.59 (0.46)
Belonged to a labor unionb 0.33 0.18 0.31 0.16
Net worth in 1993 ($1000)c 249.75 (265) 190.12 (227) 250.22 (264) 185.92 (222)
Eligible for private pension benefits
Not yet eligible 0.30 0.23 0.60 0.42
Eligible 0.41 0.24 0.11 0.06
Not covered 0.29 0.54 0.29 0.51
Occupational sector of current/last job
Private 0.64 0.68 0.66 0.71
Public 0.18 0.21 0.17 0.19
Self-employed 0.18 0.11 0.17 0.10
Occupational educationd 56.37 (27.90) 51.67 (29.59) 57.69 (27.94) 48.55 (31.30)
Duration of current employment spell 15.21 (12.55) 9.94 (9.90) 11.67 (9.78) 6.63. (7.78)
Employment status
Working full-time 0.88 0.59 0.93 0.59
Working part-time 0.08 0.14 0.04 0.20
Not working 0.04 0.27 0.03 0.22
Diagnosed with diabetesb 0.09 0.06 0.04 0.03
Diagnosed with heart diseaseb 0.06 0.02 0.02 0.01
Diagnosed with cancerb 0.05 0.07 0.02 0.03
Diagnosed with strokeb 0.01 0.01 0.00 0.01
N 4,087 4,522 95,094 103,630
Notes: Standard deviations of continuous variables are shown in parentheses.
a Observations for individuals are from the year prior to retirement or censoring.
b Proportion for whom the value is 1 = yes.
c In regression analyses, this measure is transformed to percentile rank. The mean value presented in this table is based on a variable that is topcoded at $1,000,000.
d Occupational education is a measure of occupational status equal to the percentage of persons in each occupation in the 1990 Census who completed one year
of college or more (see Hauser & Warren, 1997, for details).
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RAYMO ET AL.256
with a lower likelihood of early retirement and that belong-
ing to a labor union is positively associated with the risk of
retirement. Unlike men, however, women’s exposure to
bad jobs across the life course is unrelated to their risk of
retirement. After controlling for established correlates of
retirement timing in Model 2, the relationship between
employment in unionized jobs and earlier retirement is
Table 2. Odds Ratios from Models of Retirement Timing, by Sex
Variable
Men Women
Model 1 Model 2 Model 1 Model 2
Age
Linear 1.25** 1.18** 1.20** 1.18**
60 1.36** 1.29** 1.33** 1.25**
61 1.43** 1.46** 1.24** 1.20*
62 2.86** 2.59** 2.45** 2.15**
63 1.53** 1.46** 1.35** 1.20*
64 1.13 1.15 0.89 0.81*
65 0.93 0.94 0.69** 0.61**
Educational attainment (a)
Some college 0.94 1.02 0.92 0.92
Bachelor’s degree 0.88* 0.96 0.88 0.89
Graduate/professional degree 0.72** 0.78** 0.81* 0.81*
Marital status (b)
Divorced/separated 1.15 1.31** 0.60** 0.69**
Widowed 0.97 1.00 0.81* 0.85
Never married 1.33** 1.42** 0.91 1.03
Midlife work experiences
Involuntary job exit (c)
Before age 50 0.66** 0.84 0.89 0.95
After age 50 0.65** 0.80** 0.79** 0.86
Proportion of years w/o access to
private pension
0.49** 0.64** 0.94 0.94
Proportion of years w/o access to
health insurance
1.08 1.20 1.12 1.09
Proportion of years in low earning
occupation
0.83** 0.74** 0.91 0.86*
Belonged to a labor union (d) 1.47** 1.14** 1.35** 1.28**
Net worth in 1993 (percentile rank) 1.68** 1.96**
Eligible for private pension
benefits (e)
Eligible 3.44** 2.33**
Not covered 1.93** 1.68**
Occupational sector (f)
Public 1.59** 1.11
Self-employed 0.63** 0.82**
Duration of current employment
spell
1.01** 1.01**
Occupational education 0.99** 0.99**
Employment status (g)
Not working 1.57** 1.17**
Working part-time 1.29* 0.79*
Diagnosed with diabetes (d) 1.48** 1.23**
Diagnosed with heart disease (d) 1.13 1.72**
Diagnosed with cancer (d) 1.40** 1.27**
Diagnosed with stroke (d) 2.76** 1.34
N 95,094 95,094 103,630 103,630
df 19 32 19 32
Log-likelihood −8,921 −8,529 −10,595 −10,410
p value for likelihood ratio
test—Model 2 vs. Model 1
0.00 0.00
Notes: Omitted categories are (a) high school, (b) married, (c) none, (d) no,
(e) not yet eligible, (f) private sector, (g) working full-time.
*p < .05; ** p < .01.
Figure 1. Predicted cumulative percent retired, by age, sex, and employ-
ment experiences. Note: A refers to those with all five characteristics of interest
(i.e., experienced involuntary job loss at age 50, did not belong to a labor union,
and spent all working years employed in jobs without private pension coverage,
without health insurance, and characterized by low wages). B refers to those
who belonged to a union, did not experience involuntary job loss, and never
worked at “bad jobs.”
unchanged, but the coefficient for experience of involuntary
job loss after age 50 is no longer significantly different from
zero. Results of auxiliary models show that later retirement
among women who left a job involuntarily in late midlife
primarily reflects their relatively short employment dura-
tion at the time of retirement. One exception to the general
pattern of attenuation in the coefficients for midlife work
experiences is exposure to low-paying jobs, which is now
associated with a significantly lower risk of early retirement.
This relationship emerges after controlling for current occu-
pational socioeconomic standing which is negatively associ-
ated with exposure to low-paying jobs across the life course
as well as the risk of early retirement. In no case were these
results sensitive to the inclusion of imputed observations for
women who did not respond to the 2004 survey.
To provide a better sense of the substantive magnitude of
the estimated differences with respect to earlier employ-
ment experiences, we used predicted probabilities from
Model 2 to construct life table measures of the age-specific
cumulative proportions ever retired. For illustrative pur-
poses, we compare two extreme groups of hypothetical men
and women. One group (A) did not belong to a labor union,
experienced involuntary job loss at age 50, and spent all of
their working years in jobs without private pension or health
insurance coverage and characterized by low wages, whereas
the other group (B) belonged to a labor union, never lost a
job involuntarily, and never worked in bad jobs. Figure 1
presents the cumulative proportions ever retired by age for
these two groups of men and women, assuming mean or
modal values for other variables in Model 2 (i.e., high
school graduates; continuously married; employed in the
private sector; working full-time; not yet eligible for private
pension benefits; no health problems; and mean values of
employment spell duration, net worth, and occupational
education). The cumulative probability of retirement by age
65 for men who experienced involuntary job loss, were ex-
posed to bad jobs across the working life course, and did not
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PRECARIOUS EMPLOYMENT, BAD JOBS, LABOR UNIONS, AND EARLY RETIREMENT 257
belong to a labor union was 23 percentage points lower than
that of men without such experiences. The corresponding
difference for women was 9 percentage points. Differences
between the two groups increase with age as the age-
constant differences in the risk of retirement compound
over time. This divergence is accelerated by the experience
of involuntary job loss at the arbitrarily chosen age of 50 for
those in group A. The larger differences for men reflect the
stronger links between involuntary job loss and exposure to
bad jobs across the life course and earlier retirement for men
which more than offset the slightly stronger relationship
between labor union membership and earlier retirement for
women. Taken as a whole, these results highlight the signifi-
cance of increasingly common employment experiences
across the life course for understanding variation in the likeli-
hood of early retirement, especially among men.
Discussion
Despite the well-documented increase in exposure to
precarious employment and bad jobs and the decline in
unionized jobs, research on linkages between these em-
ployment experiences and retirement outcomes remains
limited. In this article, we used uniquely rich data col-
lected across the lives of a large cohort of older American
men and women to evaluate these relationships. We found
that experience of involuntary job exit, exposure to jobs
that do not provide pension benefits, and exposure to low-
paying jobs were all associated with a lower likelihood of
early retirement. These relationships are stronger for men
than for women. We also found a strong relationship
between labor union membership and early retirement for
both men and women.
Consistent with hypotheses based on theories of cumula-
tive stratification, these relationships were attenuated when
we controlled for established temporally proximate corre-
lates of retirement timing. Private pension eligibility and
occupational sector, in particular, played an important role
in explaining linkages between earlier employment experi-
ences and the risk of retirement through age 65. However,
employment experiences across the life course remained
significantly related to men’s risk of early retirement net of
pension eligibility, net worth, employment circumstances,
and health. Involuntary job loss after age 50, exposure to
low-paying jobs, and jobs without pension coverage re-
mained significantly associated with a lower risk of early
retirement, whereas labor union membership continued to
be positively associated with early retirement. For women,
exposure to low-paying jobs emerged as a significant cor-
relate of later retirement, whereas union membership re-
mained a strong predictor of early retirement. In sum, we
find that exposure to precarious employment and bad jobs
are associated with a lower likelihood of early retirement,
especially for men. Part, but not all, of these relationships
can be attributed to the fact that those with more exposure to
precarious employment and bad jobs are less likely to be
eligible for private pension benefits, more likely to be self-
employed, and have shorter employment duration as they
approach typical retirement ages.
Gender differences in the meaning of work and career
may account for the relatively weaker linkages between
women’s earlier work experiences and their retirement
timing. The WLS respondents belong to a transitional co-
hort between earlier cohorts of women who derived their
identity primarily from family roles and later cohorts who
have derived their identity from both work and family roles
(Goldin, 2006). We suspect that women’s retirement timing
may be more sensitive to their husbands’ characteristics
than to their own employment history, but the absence of
time-varying information on spouses’ characteristics
prevents us from examining this possibility. It will be im-
portant to reexamine gender differences in younger cohorts
characterized by greater symmetry in men’s and women’s
employment experiences across the life course when
appropriate data are available. It will also be important to
pay attention to other characteristics that may moderate re-
lationships between employment experiences across the life
course and retirement timing. For example, labor union
membership and self-employment are associated with em-
ployment stability and job quality in ways that may shape
relationships between employment experiences across the
life course and retirement outcomes.
Our results are particularly interesting in the context of
increasing interest in promoting extended labor force
attachment at older ages, suggesting that exposure to less
favorable employment circumstances across the life course
may contribute to a lower likelihood of early retirement.
Results are consistent with the hypothesis that changing
employment patterns across the life course contribute to in-
creasing heterogeneity in the timing of retirement via the
rising proportion of older Americans who approach older
ages without sufficient financial resources and benefits to
retire early (Moore & Mitchell, 2000). However, our ability
to do more than speculate about this possibility is con-
strained by two important limitations of our study.
The first limitation is that our ability to adequately mea-
sure established temporally proximate correlates of retire-
ment timing is constrained by the information available in
the survey. Although the WLS does contain information
about savings, financial assets, and pension benefit levels,
these measures refer to the time of the survey and are thus
not adequate for constructing the truly time-varying indica-
tors of economic well-being needed for our analyses. This is
particularly true of pension benefit levels, which were
assessed only in the 2004 survey and thus cannot be mean-
ingfully incorporated in analyses of retirement across the
full range of ages that we examine. Careful evaluation of
the mechanisms underlying the relationships we have
documented is an important task for subsequent research
when the necessary detailed information on individual
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RAYMO ET AL.258
employment, economic, and health characteristics across
the life course become available. The first large-scale study
to provide such information will be the NLSY79, whose
participants will enter prime retirement ages 10–15 years
from now.
A second limitation is the fact that the WLS contains only
limited information about the life histories of spouses.
Spouses’ employment experiences across the life course
and their economic, employment, and health circumstances
at older ages may be related to respondents’ midlife
employment experiences and retirement timing in ways that
would shed light on our findings. For example, the rela-
tively low likelihood of early retirement among men with
less favorable employment circumstances in midlife may
reflect their marriage to women with similar employment
experiences or perhaps more limited labor force attachment
across the life course. We also recognize that exposure to
precarious employment and bad jobs across the life course
may affect retirement timing in ways that have yet to be
adequately theorized. For example, it may be that exposure
to precarious employment and bad jobs across the life
course influences preferences for work and leisure, life goals,
risk aversion, and other characteristics that may affect re-
tirement timing but are not included in the WLS or in most
other surveys.
Refining analyses in these ways to further our under-
standing of linkages between employment experiences
across the life course and retirement timing is of potentially
great value for understanding variation in the retirement
process of the large baby boom cohort. Relative to the WLS
cohort and preceding cohorts, the baby boomers were at an
earlier career stage when marked changes in the nature of
employment emerged in the mid-1970s. A better under-
standing of the ways in which earlier employment experi-
ences are associated with retirement timing is valuable not
only for social scientists but also for individuals contem-
plating and planning for their own retirement and for those
involved in the formulation of retirement policies. Social
scientists have discussed and documented the deinstitution-
alization or individualization of retirement, but they have
yet to fully explore the ways in which employment experi-
ences across the life course influence retirement outcomes
in this new context. As more individuals approach retire-
ment with plans for extended employment, including part-
time work or phased retirement, effective planning and
preparation may be enhanced by a fuller understanding of
the ways in which earlier employment experiences are as-
sociated with the likelihood of early retirement. Similarly,
the efforts of individual firms and policy makers to imple-
ment policies designed to facilitate extended labor force
participation will benefit from an understanding not only of
the prevalence and nature of plans for work at older ages,
but also of the ways in which individual variation in earlier
employment experiences contribute to variation in the risk
of early retirement.
Funding
The research reported herein was supported by the National Institute on
Aging (R01 AG-9775 and P01-AG21079), by the National Science Foun-
dation (NSF-0550752), by the William Vilas Estate Trust, and by the
College of Letters and Science and the Graduate School of the University
of Wisconsin–Madison. Additional support was provided by the Institute
of Social and Economic Research at Osaka University. Research was
conducted at the Center for Demography and Ecology and the Center for
Demography of Health and Aging at the University of Wisconsin–Madison,
which are supported by Center Grants from the National Institute of Child
Health and Human Development (R24 HD047873) and the National In-
stitute on Aging (P30 AG17266). The opinions expressed herein are those
of the authors. Data from the Wisconsin Longitudinal Study used in this
analysis are publicly available at http://www.ssc.wisc.edu/wlsresearch/.
Correspondence
Correspondence should be addressed to James M. Raymo, PhD,
Department of Sociology, University of Wisconsin, 1180 Observatory
Dr., Madison, WI 53706. E-mail: jraymo@ssc.wisc.edu.
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