A

tpswc©

K

aadsdtmddwatn

Ic

0

Research in Social Stratification and Mobility 24 (2006) 223–238

Widening the gap: The effect of declining unionization onmanagerial and worker pay, 1983–2000�

Jake Rosenfeld ∗Department of Sociology, Office of Population Research, 284 Wallace Hall, Princeton University,

Princeton, NJ 08544-2091, United States

Received 14 April 2005; received in revised form 17 November 2005; accepted 20 January 2006

bstract

This paper examines the relationship between falling union membership and the pay gap between workers and managers duringhe 1980s and 1990s. Analysis of industry-level data from the Current Population Survey indicates that unions boost median worker

ay and are also associated with slightly higher mid-level managerial pay, as the union wage premium reverberates up the paycale. Despite the positive association with both median worker and managerial pay, estimates indicate that union decline widensage dispersion within the workplace. The wage premium unions offer workers dwarfs the positive union effect on managerial

ompensation, suggesting that unions operate to influence the underlying pay norms of a firm in the wage determination process.2006 Elsevier Ltd. All rights reserved.

eywords: Union membership; Industry-level data; Income inequality

During the past few decades of growth in executivend top-level managerial compensation, real wages forverage workers in the United States have stagnated oreclined (see Morris & Western, 1999 for a comprehen-ive review of the literature). Wages for the middle twoeciles of the income distribution remained flat duringhe 1980s only to fall precipitously during the early and

id-1990s. Indeed, average real wages for nearly everyecile of the American income distribution stagnated orropped during the final decades of the 20th Century,ith the marked exception of the top 10%. As Morris

nd Western conclude, “the story of this period is nothat the rich got richer and the poor got poorer, but thatearly everyone lost ground” (1999: 626).

� A previous version of this paper was presented in the sessionnequality at Work at the annual meetings of the Eastern Sociologi-al Society, New York, New York, February 19–22, 2004.∗ Tel.: +1 609 258 5508.

E-mail address: jhrosenf@princeton.edu.

276-5624/$ – see front matter © 2006 Elsevier Ltd. All rights reserved.doi:10.1016/j.rssm.2006.01.003

The influence of unions on wage-setting fell substan-tially both in the U.S. and abroad during this periodof growing income polarization (Western & Healy,1999). Many researchers have documented the signifi-cant drop in private-sector unionization levels beginningin the 1970s (an excellent summary of this phenomenonis provided by Clawson & Clawson, 1999; see alsoCard, 1998). Between the mid-1970s and mid-1990s,the private-sector unionization rate among men declinedby nearly 50% in the U.S. (Card, 1998). The 1980swere especially harsh on organized labor: in industries asvaried as transportation, paper production and primarymetals, union representation fell by nearly 25% in lessthan 10 years.

Past research on the relationship between uniondecline and pay inequality often focused on the impact

of unions on non-professional, non-managerial work-ers. Researchers specified two main mechanisms throughwhich unions boost worker pay. One is the union wagepremium, which refers to the increased pay unions are

atificati

224 J. Rosenfeld / Research in Social Str

able to provide their members directly through collectivebargaining (see Lewis, 1986). Secondly, numerous anal-yses have found union threat effects in certain regionsand industries (Corneo & Lucifora, 1997; Leicht, 1989).Non-union employers, worried about the threat of labororganizing, attempt to preempt unionization by raisingwages to union levels.

The well-documented wage premium for unionizedworkers and the pay raise due to the threats of unioniza-tion may not be the only effects an organized workerpresence has on workplace wage inequality. Sociolo-gists describe cases in which unions mobilize to reduceinequities that go beyond the narrow wage interests oftheir members. For example, CIO unions in the 1930sand 1940s pursued a broad agenda of economic equalityby pushing for greater gender and racial equality in theworkplace (Milkman, 1987; Stephan-Norris & Zeitlin,2003). If unions care about reducing inequality, thenunion influence on the distribution of pay may extendbeyond the wage premium exacted for the organization’sown members.

This paper studies the effects of union decline on man-ager to worker pay inequality between 1983 and 2000.I measure inequality with the ratio of median manage-rial to median non-professional worker pay in specificindustries and regions.

Changes over time in the distance between manage-rial and worker pay in particular industries and regionscan provide some indication of the normative influenceof unions on workplace pay. If those workplaces witha highly unionized presence have a comparatively com-pressed pay schedule all the way up the managerial line,then it is evident that not only do unions offer their mem-bers a pay boost, but they also exert some pressure onmanagement for a more equitable pay distribution.

First I examine whether the ratio between medianmanagerial pay and median worker pay has increased inrecent decades, and what various industry and regionalfactors might explain the trends. I then investigatewhether the drop in private-sector unionization explainsany time trend independent of various other labor marketchanges. Finally, I conduct analyses on managerial payand worker pay separately to see what exactly explainschanges in the ratio of the two. For all investigations, theunit of analysis is the industry-region-year.

1. Unions, worker pay and managerialcompensation

The importance of occupations for studying socialstratification often distinguishes sociological researchfrom work in economics. Sorenson (1996), for exam-

on and Mobility 24 (2006) 223–238

ple, argues that certain economic rewards, or rents,attach to particular occupations regardless of the per-sonal characteristics of the occupants of the position.Sociologists have emphasized the primacy of occupa-tions in rent seeking within firms (Grusky & Sorenson,1998; see also Sorenson, 1996). If rents attach to occupa-tions, then those occupations with organizations able tocapture rents – like unions – should succeed in reward-ing its members relative to occupations lacking such anorganizational base. Moreover, if rent-seeking occurs atthe occupational level, then union effects on pay normsshould be felt at this level of disaggregation. The ratioof managerial to worker pay measured within particularindustry-region cells captures how the institutionalizedpresence of unions influences the distribution of rents todifferent occupational groupings: specifically, workersand their managers.

The precipitous decline in union membership duringthe past decades corresponds with a sharp increase ininter-occupational pay inequality (see Fig. 1). By theend of the 1990s, union membership levels settled ataround 12% of the full-time non-professional workforce.Meanwhile, the ratio between median managerial payand median worker pay – despite small fluctuations –increased markedly between 1983 and 2000, with themost rapid climbs occurring in the late 1990s.

Declining unionization rates and increased manager–worker inequality might be related in several ways.Unionization may increase median worker wages rela-tive to salaries at the top of the pay scale, and this increasein worker pay drives down the ratio, or unionizationdrives down median managerial pay, thereby reducingthe ratio, or both. I briefly explore the theoretical ratio-nale underlying these possible scenarios below.

1.1. Relationship between unionization andworker pay

Labor market researchers have consistently demon-strated that unionization increases average wages amongotherwise comparable workers (Lewis, 1986; see Kuhn,1998 for an overview of the more recent literature).Cross-sectional analyses reveal that among observa-tionally equivalent workers, unionization boosts wagesaround 15% (Robinson, 1989). Various analyses usingpanel data have largely corroborated this 15% find-ing (Card, 1996; Freeman & Medoff, 1984; Kuhn &Sweetman, 1998). Other research on the topic suggests

that for recent years – due to match bias and measure-ment error in the Current Population Survey – the 15%boost is actually too low (Hirsch, 2004). How do unionsachieve such impressive wage gains for their members?

J. Rosenfeld / Research in Social Stratification and Mobility 24 (2006) 223–238 225

F pay ratii rreprese

Faisptcb1maunW

sifipwUr1aediw

ig. 1. Trends in unionization levels and managerial/worker weeklyndustry-region-year groupings with relatively few entries are not ove

rom an economic perspective, unions are able to capturewage premium for members in concentrated industries

n which higher labor costs can be passed on to con-umers (Rees, 1989). Sociologists have emphasized theotentially disruptive location of union employment inhe production chain: in capital-intensive industries likeoal and auto, labor’s “positional power” enables it toargain for higher wages (Wallace, Griffin, & Rubin,989; see also Perrone, 1983, 1984). Even union employ-ent in non-strategic locations may obtain higher wages,

s the organization of workers into collective bargainingnits presents a unified front during times of contractegotiation and organizational restructuring (Kalleberg,allace, & Althauser, 1981).The 15% wage boost attributable to union member-

hip probably understates the role unions play in affect-ng worker pay. Evidence suggests that the threat ofurther union organization and the pay inequities result-ng from the union wage premium cause employers toay higher wages to their non-union workers in firmsith some union presence, ranging from 43 cents toS$ 3.89 for every dollar increase in the union wage

ate (Leicht, 1989, p. 1042; see also Freeman & Medoff,984). In their study of Italian engineering firms, Corneond Lucifora (1997) find threat effects especially pow-

rful in establishments with intermediate levels of unionensity. Neumark and Wachter (1995) find that at thendustry-level increases in the percentage of organizedorkers may depress comparable non-union pay. At the

o, 1983–2000. I weight calculations by cell frequency, ensuring thatnted in the figures.

city-level, however, the authors find the opposite effect:a highly organized workforce tends to raise wages forall workers, suggesting that threat effects predominateat this sub-regional level. Other research has establishedthe presence of interindustry threat effects in firms wherenon-unionized workers are linked in the productionchain to heavily unionized industries (Leicht, Wallace,& Grant, 1993; Martin & Rence, 1984).

1.2. Relationship between unionization andmanagerial pay

While researchers have studied union effects on thepay of unionized workers, or comparable non-unionworkers, few have looked at the effect of unions on man-agerial compensation. I argue that unions may help estab-lish pay norms that govern workers and their managers.Contrary to the economic analysis of union rent-seeking,my account suggests that organized labor is not sim-ply a self-interested actor intent on maximizing wagesand benefits for its own members. Describing egalitar-ian collective bargaining in Sweden, Swenson (1989, p.3) argues that in order to cultivate and maintain internaland external support, unions inhabit a “moral economy”where they attempt to shape the values that underlie the

entire wage structure. Beyond simply extracting wagegains for their members, unions seek to instill wagenorms that narrow the firm’s pay distribution beyondwhat purely market forces would dictate. This idea of

atificati

226 J. Rosenfeld / Research in Social Str

a union’s moral economy may not be confined to Scan-dinavia. In their analysis of radical CIO unions duringthe 1930–1950s, Stephan-Norris and Zeitlin (2003) doc-ument union efforts to expand gender and racial equal-ity (see also Milkman, 1987; Zeitlin & Weyher, 2001).Direct wage gains for their members may not be the onlyconcern for unions within the pay determination processeither.

There are also examples from contemporary laborrelations in which executives of unionized firms are lessinclined to grant themselves significant pay increases.Freeman and Medoff (1984) report that after GeneralMotors substantially increased executive pay in 1982, theensuing uproar from the highly organized plant workersforced the bosses to rescind their own raises. Moreover,in periods of concession bargaining (increasingly com-mon through the 1980s and 1990s), unionized employeesmight demand freezes on managerial pay to match theirown (Mitchell, 1994). For example, after airline unionsagreed to deep wage cuts in an effort to avoid bankruptcyin the spring of 2003, union pressure helped force theresignation of American Airlines CEO Donald J. Cartywhen it was revealed that he had negotiated bonuses forhimself and other top executives (Wong, 2003). Suchoccurrences hint at how an organized worker presencecan have a normative effect on workplace pay scales.

Unions might reduce the gap in pay between workersand managers in several ways. They may lower the payof managers while raising the pay of unionized workers,thereby reducing the wage gap between the two. Sucha finding could indicate unions’ ability to influence paynorms within the workplace (see Swenson, 1989 for adiscussion of pay norms). Or the negative union effect onmanagerial pay could simply reflect the lower rents avail-able to managers after unionized workers have extractedtheir wage premium. On the other hand, unions mayoperate to lower workplace pay inequality by simulta-neously boosting average worker and average managerpay, but at differing levels (i.e., by granting workers adisproportionately higher pay raise than their managers).Freeman and Medoff (1984, pp. 174–180) suggest that incertain firms in certain industries unions operate to boostproductivity. These productivity increases may translateinto higher take-home pay for all employees—includingmanagers. However, if the managerial pay boost doesnot equal the combined effect of the union wage pre-mium and the increased worker pay due to productivityincreases, the gap between worker and managerial pay

will still shrink.

Such a scenario renders the rents explanation less ten-able. After all, the lower rents hypothesis suggests thatthe pay boost unions offer workers reduces the amount

on and Mobility 24 (2006) 223–238

of money left over for the rest of the pay scale, therebydriving managerial compensation downward. The paynorm hypothesis still holds if, in fact, a union presenceincreases managerial pay. An organized worker presencemay influence wage-setting norms by decreasing the dis-tance between the pay of average workers and that oftheir bosses, while providing both sets of workers witha wage boost.

2. Prior research

Research on the role of union decline in wideningwage inequality falls into two general categories. Thelargest focuses on organized labor’s role in increasingwages for non-professional, non-managerial workers.The second, smaller body of literature attempts to cap-ture union effects on managerial pay.

2.1. Union decline and worker pay inequality

Researchers have consistently attributed a significantportion of the recent rise in income inequality to labor’sdecline (DiNardo, Fortin, & Lemieux, 1996; Freeman,1993). In the traditional account, unions equalize wagesby raising the pay of those at the bottom of the wagedistribution, either through the union wage premiumor threat effects. Card’s findings (1998) indicate thatbetween 10% and 20% of the recent growth in male earn-ings inequality is due to falling union membership. Com-paring wage inequality in Canada and the United Statesduring the 1980s, DiNardo and Lemieux (1997) con-clude that fully two-thirds of the differential in earningsinequality between the two nations is due to America’ssteeper union decline. In their historical evaluation ofwage inequality, Goldin and Katz (1999) find that peri-ods of wage narrowing are almost always associated withincreased union activity—and the historically unprece-dented rate of union loss during the past few decadeshelps to explain our current period of growing incomepolarization.

2.2. Effects of declining unionization on managerialpay

Recently, a few studies have investigated the relation-ship between unionization and managerial pay. In a studyof Canadian metal-mining firms, Singh and Agarwal(2002) find that union presence is associated with higher

CEO pay, although the relationship disappears whencontrolling for firm size and performance. DeAngeloand DeAngelo (1991) investigate the effects of workerorganization on CEO pay in the American steel indus-

atificati

ttydn

Heapyadpmatt2AsbtaFa

mtwbbdawd(n

3

asedaTmTvr

and an industry-region measure of union contract cov-erage. The two items do not vary dramatically from oneanother by industry, region or year, but the union cov-

1 I also re-ran the main models excluding managerial cell counts that

J. Rosenfeld / Research in Social Str

ry during the 1980s, and conclude that unions operatedo reduce executive pay by 18.1% in union negotiationears. The authors suggest that executives take pay cutsuring periods of industry contraction to persuade unionegotiators that worker wage concessions are necessary.

In the most comprehensive test thus far, DiNardo,allock, and Pischke (1997) do not find significant

ffects of unionization on managerial pay, but do findgeneral negative correlation between executive com-

ensation and unionization in their cross-sectional anal-ses. Moreover, in their analysis of union elections, theuthors conclude that a loss of members due to unionecertification results in 10–15% higher executive com-ensation for the CEOs of newly non-union firms. Inore recent work, the authors speculate that lower wages

mong managers in highly unionized industries reflectshe redistribution of rents from supervisory positions tohe workers themselves (DiNardo, Hallock, & Pischke,000). How might unions achieve such redistribution?s a collective voice for workers, organized labor may

uccessfully lobby for wage increases during collectiveargaining, thereby reducing the available rents. Strikehreats or strikes themselves may frighten managementway from granting themselves large pay increases.inally, management itself might use top-level pay cutss a bargaining tactic against union wage demands.

To date, no research evaluates union effects on bothanagerial and worker pay simultaneously. It may be

hat unions depress managerial pay while increasingorker pay in union firms. Or unionization may raiseoth average workers’ pay and the pay of their managers,ut at differing levels. Such situations would provide evi-ence that unions concern themselves not only with thebsolute wage levels of their members, but with theirage levels in relation to others in the firm’s occupationalistribution. Such scenarios would buttress Swenson’s1989) claim that unions seek to shape the underlyingorms governing a firm’s pay distribution.

. Data and methods

Earnings and union data are taken from the CPSnnual merged outgoing rotation file, 1983–2000. Myample is restricted to non-professional private-sectormployees working full-time with sufficient earningsata. I exclude the self-employed. Following DiNardo etl. (1997), I divide occupations into two broad groups.he first group encompasses executives and top-level

anagers (1980 three-digit SIC occupation codes 3–22).he primary responsibility of these workers is to super-ise others, and under National Labor Relations Boardegulations they are prohibited from organizing into

on and Mobility 24 (2006) 223–238 227

unions. A small number of executives and managers indi-cate that they belong to a union and these respondents aredropped from the analysis. The second group includesall non-professional, non-managerial full-time workers.I calculate industry-level unionization rates based on thisgroup’s membership status.

Since my primary analysis is at the industry-region-year level, I aggregate each year of CPS individual leveldata into a final dataset with cell entries for each industry-region-year grouping. To maintain adequate cell sizes,I aggregate industry codings into 26 relatively broadgroupings (DiNardo et al., 1997; detailed industry listprovided in Appendix A). To control for regional effects,industry location is divided into four major regions (stategroupings provided in Appendix B). My final samplesize is 1872 (4 regions by 26 industries by 18 years).For workers, cell counts never fall below 30, but forcertain industries (forest/fisheries, for example) manage-rial cell counts are quite low in some of the industry-region-year cells. To correct for these small cell counts,all models are weighted by managerial cell count size,although the weighting procedure makes little substan-tive difference.1

The analysis examines several managerial and non-professional worker pay variables for each industry-region-year cell, using the appropriate CPS weights. Formy primary analysis (Table 2), I utilize a measure cap-turing the ratio of median managerial to median weeklyworker pay. Median weekly pay measures for both work-ers and managers avoid topcoding and outlier bias thatmight pose a problem if I had instead chosen mean mea-sures. Given that some of my sample’s cells are quitesmall, median pay items provide the most robust esti-mate for average wages (Wright & Dwyer, 2003 usesimilar CPS wage measures). For Table 5, I examineunionization effects on 90th percentile managerial pay.Given small cell sizes as well as CPS topcoding proce-dures, the results from Table 5 should be interpreted aspreliminary. Wage measures throughout the analyses areexpressed in constant US$ 2000.2

The independent variables of primary interest includean industry-region measure of workforce unionization

fell below 18 (the 10% cutoff in the managerial cell size distribution)and 39.5 (the 25% cutoff). Results do not differ substantively from theones in the models shown and are available upon request.

2 Analyses were also conducted using log median wage measures,with similar results.

atification and Mobility 24 (2006) 223–238

Table 1Weighted means of workforce demographic and wage measures

1983 1992 2000

Manager/worker pay ratioa 1.77 1.81 1.90

Non-professional workersPercent unionized 22 15 12Percent female 42 44 43Percent minorityb 17 21 28Percent high school degree 69 44 41Percent at least some college 10 39 43Median weekly wage (US$ 2000) 474 465 481

Managers/executivesPercent female 31 40 42Percent minorityb 5 8 12Percent some college 10 20 20Percent bachelor’s degree or higher 61 54 58Median weekly wage (US$ 2000) 834 838 909

a These manager/worker pay ratios reflect actual points in the data,

1980s.4 The managerial ranks saw substantial increasesin both minority and female representation during thetime period under investigation.

228 J. Rosenfeld / Research in Social Str

erage measure captures some non-union members whononetheless enjoy the benefits of working under a unioncontract. I run all models with both union measures sep-arately; results from the analyses do not differ by choiceof measure.

Covariates that could affect changes in pay over timeinclude a range of demographic characteristics, such asthe racial makeup of the managers and workers, gendercomposition, potential experience and educationallevels. Human capital theory posits that increasingeducation levels should translate to wage gains forboth workers and managers. Controlling for levelsof education, the relegation to lower-paying taskswithin occupations should depress median pay inhigh-minority or high-female cells. Controlling for theworkforce characteristics is also necessary since unionmembership is spread unevenly across the demographiclandscape: more men belong to unions than women,and historically minorities and the lower-educatedcomprised a disproportionate share of the union rolls.Moreover, evidence suggests that unions affect groupsdifferently. The union wage premium benefits lowereducated workers disproportionately (although the gap isnarrowing), as well as non-whites compared with whites(Blanchflower & Bryson, 2003; see also Freeman &Medoff, 1984).

To control for these demographic and workforce com-position shifts in employment that may influence man-agerial and worker pay scales, I have items capturing thepercentage of non-professional workers that are femalefor each industry-region grouping, the percentage ofmanagers who are black, the percentage Hispanic, aver-age managerial educational levels (including the frac-tion that completed some college, the fraction with abachelor’s degree or higher and the fraction with no col-lege experience) and average ages. The dataset includescorresponding variables for managers for each industry-region cell.3 As I discuss below, my choice of modellargely captures any unmeasured industry or regionallyspecific features that affect pay scales, such as averageproductivity levels.

Table 1 presents aggregate descriptive statistics forvarious manager and worker demographic measures,

weighted by cell frequency. Union participation ratesnearly halved during the time period under investigation.Workplace inequality, captured by the ratio of median

3 Although given the different educational distributions of the twoclasses of workers, the education variables for workers differ frommanagers: the education items for workers divides the sample into thosewho never completed high school, those with a high school degree andthose with some college experience.

not measures predicted by the models used in the subsequent analysis.b Minority category restricted to self-identified African–Americans

and Hispanics.

managerial to median worker pay, increased around 7%during the 1980s and 1990s. Unsurprisingly, the meanlevel of education for workers has increased sharply inrecent years. Despite the increasing educational levelsof these workers, median wages remained quite stag-nant, leveling off at the end of the twentieth centuryonly 1.5% higher than where they stood in the early1980s.

Median managerial weekly wages (expressed in con-stant US$ 2000) rose substantially over the time period,although the bulk of this wage growth seems restrictedto the 1990s. Unlike workers, where one sees a real risein education levels, the educational distribution of man-agers remained rather stagnant throughout the 1980sand 1990s. By 2000, over three quarters of managersreported at least some college experience, a percent-age only slightly higher than where it stood in the early

4 Due to changes in CPS education items over survey years, educa-tion scores represent rough approximations of respondent’s educationand not perfect measures. Beginning in 1992, the CPS replaced theiryears of schooling measure with a credential measure. This leadsto some slippage in comparing educational outcomes between theyears, especially at the higher end of the educational distribution. Forinstance, given that time to completion for a Bachelor’s degree canrange quite considerably (from three to many, many years), one can-not determine with precision those who graduated from college in thepre-1992 sample.

atificati

dw

piab

y

wfiateoiiusafinttwatrttda

n1Si(uhaeaeppepti

wage boost unionization offers non-professional, non-managerial workers. According to the model, a one-pointincrease in unionization is associated with a US$ 3.93

J. Rosenfeld / Research in Social Str

Both Table 1 and Fig. 1 depict the basic trend ofeclining unionization and a growth in managerial-orker inequality. How are these two trends related?To best estimate the impact of unionization on occu-

ational pay inequality, I fit a fixed-effects model thatncludes a separate effect for each industry-region andtime effect for each period (in this case, year rangingetween 1983 and 2000). For industry i at time t,

it = αi + νt + x′itβ + εit

here αi is the fixed industry-region effect, νt thexed year effect, x′

it represents the exogenous vari-bles in industry-region i at year t and εit is the errorerm. The fixed-effects control for unmeasured differ-nces between industry-regions that remain constantver time, and control for forces that affect the var-ous industry-regions equally but vary over time. Fornstance, the fixed industry-region effects control fornmeasured industrial characteristics like average firmize, industrial concentration or the lower wage scalesttributable to regional differences, whereas the yearxed-effects capture the influence of a macro-level eco-omic event like the recession of the early 1990s. Givenhat historically unions concentrate in monopoly sec-ors where employers can more easily pass on the unionage premium to consumers, and where employers can

fford higher wage demands of rent-seeking managers,he fixed-effects help determine whether higher wageseflect unions’ presence or simply higher rents. Whilehis investigation does not completely solve the problem,he industry-region and year fixed-effects lend confi-ence to drawing a causal explanation between unionsnd wages.

This quantitative analysis builds upon labor eco-omics research (specifically DeAngelo & DeAngelo,991; DiNardo et al., 2000; Neumark & Wachter, 1995;ingh & Agarwal, 2002), and expands upon the exist-

ng literature in significant ways. While DiNardo et al.2000) used a similar analysis to capture the effects ofnion presence on managerial pay scales, no other studyas specifically examined the relationship between man-gerial pay and average worker pay, and what mightxplain changes in the gap between the two. This setupllows me to simultaneously focus on two hypothesizedffects of unionization on wages: one, the demonstrateday boost attributable to union membership; two, the

ressures an organized worker presence exerts on thentire pay distribution of a workplace. My focus on theivotal decades of the 1980s and 1990s perfectly targetshe period of rapid of de-unionization and growing wagenequality.

on and Mobility 24 (2006) 223–238 229

4. Results

Table 2 shows the results of a series of models predict-ing changes in the ratio of managerial wages to workerwages over time. The effect of the union variable remainsrobust across various model specifications. The simplestmodel (Model 1) includes a year variable to capture anylinear time trend in the managerial/worker pay ratio andseveral workplace demographic measures. The unioncoefficient is negative and significant at the .001 level. InModel 2, I introduce a year fixed-effect, controlling forany non-linear time trend that might affect pay inequal-ity between managers and workers during the 1980sand 1990s. The year fixed-effect boosts the amount ofvariation captured by the model; despite this furtherspecification, the union coefficient remains significantand negative. Demographic measures are all significantand signed in the expected directions.5

Model 3 includes an industry-region fixed-effect, con-trolling for the impact of any unmeasured variables thatvary across industry-regions. Model 3 captures over two-thirds of the variation in the estimate of the manage-rial/worker pay ratio. The union coefficient of −.610indicates that a 10 point drop in the percent unionizedis associated with about a .06 point increase in the ratioof worker to managerial pay, or nearly half of the totalincrease in pay inequality during the 1980s and 1990s.

Table 3 attempts to determine which component ofcross-occupational pay inequality is affected by changesin unionization levels. It could be that declining union-ization has no effect on median managerial pay, and sim-ply works to boost median worker pay, and thus reducethe ratio between the two. Or the data could reveal thatunionization levels do not affect worker pay (a highlyunlikely scenario), but drag down average managerialcompensation, thereby decreasing the ratio. Or the datamay reveal a combination of forces at work.

I first investigate the relationship between unionmembership and worker pay. The dependent variable inModel 1 of Table 3 is median weekly wages for private-sector, non-professional, non-managerial workers. Theunion coefficient is positive and significant at the .001level, revealing that at least part of the impact of union-ization level on the managerial to worker pay ratio is the

5 I ran all models with the minority variable broken out into percentHispanic and percent African–American. Results do not differ sub-stantively from the models displayed in the paper, and are availableupon request.

230 J. Rosenfeld / Research in Social Stratification and Mobility 24 (2006) 223–238

Table 2Fixed-effects models predicting workplace pay inequality

Model 1 Model 2 Model 3

Proportion unionized −.198*** (.048) −.299*** (.050) −.610*** (.151)Year .025*** (.002) – –

Worker variables% Female 1.303*** (.043) 1.293*** (.040) .749*** (.161)% Minority .179* (.078) .186* (.078) .547** (.168)HS degree −1.533*** (.090) −1.386*** (.093) −.855*** (.187)College −1.595*** (.079) −1.935*** (.081) −1.106*** (.219)Age −.119 (.063) −.259*** (.063) −.144* (.068)Age2 .001 (.001) .003*** (.001) .002 (.001)

Manager variables% Female −1.582*** (.048) −1.431*** (.049) −.692*** (.070)% Minority −.396*** (.108) −.334** (.110) −.227* (.110)Some college .525*** (.106) .463** (.102) .082 (.103)College degree 1.426*** (.056) 1.323*** (.056) .633*** (.077)Age −.113** (.035) −.077* (.032) −.030 (.040)Age2 .001** (.000) .001** (.000) .001 (.000)

Year fixed-effects No Yes YesIndustry/region fixed-effects No No YesConstant −42.759*** (3.314) 8.552*** (.994) 5.202*** (1.258)N 1872 1872 1872Number of parameters 15 31 134R2 .67 .68 .79

Note: Standard errors are in parentheses. Dependent variable in all models is the ratio of median managerial to median worker weekly pay. Wagesin constant US$ 2000. Models weighted by the number of managers in each cell.

* p < .05.

** p < .01.

*** p < .001.

raise in median weekly wages. Other worker covariatesin Model 2 trend in the expected directions.

Fig. 2 displays the estimated trajectory of worker payif unionization remained at its 1983 levels using the esti-mates from Table 3. Annual results are weighted by thenumber of workers in each industry-region cell. Whilethe data point for each individual year should be treatedwith some caution due to relative small sample sizes (104for each year) and large standard deviations, the overalltrend is clear: on average, if unionization levels had notdecreased from 1983 on, median weekly worker pay isestimated to be around 3% higher.

Next, I test to see what effect declining unionizationhas on median managerial pay. The dependent variablein the second model of Table 3 is median weekly man-agerial wages. Somewhat surprisingly, the union coef-ficient remains significant—and positive. According tothe model, a one-point rise in the percentage of workers

unionized is associated with a US$ 3.13 increase in mid-level managerial pay. For supervisors in the middle of themanagerial pay distribution, a relatively high union pres-ence works to boost wages. High union presence pushes

wages up at the worker level and reverberates through-out much of the pay scale. This result provides evidenceagainst the hypothesis that unions act to restrain mid-level managerial pay in the workplace.

High levels of union membership at the industry-region level actually seem to drive both median man-agerial pay and median worker pay upward. How doesthis finding square with the negative effect of unions onoccupational pay inequality? When analyzed separately,higher levels of union membership positively correlatewith both median managerial pay and median workerpay. However, when looking at the ratio of the two, itis clear that the positive effects of high levels of mem-bership on worker pay overwhelm the smaller effectson managerial pay, and therefore the models of Table 2reveal a significant negative relationship between unionlevels and cross-occupational weekly pay inequality.

These findings are corroborated in Fig. 3 below, where

once again I fix unionization at its 1983 level for allindustry-region cells. This time, I predict weekly man-agerial wages, weighting the yearly results by the numberof managers in each cell. As shown, higher levels of

J. Rosenfeld / Research in Social Stratification and Mobility 24 (2006) 223–238 231

Table 3Disaggregating the effect of unionization on workplace pay inequality

Worker pay Managerial pay

Proportion unionized 392.541*** (18.933) 313.371*** (68.004)

Worker variables% Female −131.735*** (20.121) 115.984 (72.271)% Minority −158.092*** (21.084) −30.008 (75.728)HS degree 174.815*** (23.377) −19.380 (83.965)College 246.062*** (27.391) 18.304 (98.383)Age 25.571** (8.495) 1.067 (30.512)Age2 −.306** (.113) −.044

Manager variables% Female .194*** (8.779) −304.093*** (31.534)% Minority −27.216* (13.816) −159.194** (49.624)Some college −17.711 (12.861) 23.078 (46.194)College degree 38.581*** (9.581) 374.852*** (34.412)Age 27.056*** (4.978) 15.407 (17.881)Age2 −.342*** (.062) −.086 (.221)

Year fixed-effects Yes YesIndustry/region fixed-effects Yes YesConstant −788.743*** (157.421) 75.868 (565.419)N 1872 1872Number of parameters 134 134R2 .97 .92

Note: Standard errors are in parentheses. Model 1 dependent variable: median weekly worker pay. Model 2 dependent variable: median weeklymanagerial pay. Wages in constant US$ 2000. Models weighted by the number of managers in each cell.

* p < .05.** p < .01.

*** p < .001.

Fig. 2. Predicted median worker wages based on actual and 1983 unionization levels. I weight calculations by cell frequency, ensuring that industry-region-year groupings with relatively few entries are not overrepresented in the figures.

232 J. Rosenfeld / Research in Social Stratification and Mobility 24 (2006) 223–238

3 unionrreprese

Fig. 3. Predicted median managerial wages based on actual and 198industry-region-year groupings with relatively few entries are not ove

unionization correlate with slightly higher median man-agerial pay. According to the model, if unionizationlevels remained at where they were in 1983, medianweekly managerial pay would be about 1.5% higher.Since union presence actually works to increase man-agerial as well as worker pay, labor’s overall effecton the pay inequality ratio is muted: if unionizationremained at 1983 levels, pay inequality between work-ers and their managers would be about .036 pointslower.

How does the effect of union decline compare withother variables? In Table 4, I calculate the overall effectof unionization on the pay inequality ratio, workerpay and managerial pay after fixing the percent union-ized at its 1983 level. I construct similar figures fortwo of the most powerful covariates in the full model,

the percent female managers and the percent minorityworkers. The results show that had the percentage offemale managers not risen above its 1983 level, man-agers would have, on average, enjoyed a US$ 20 increase

Table 4Predicted effects of covariates fixed at 1983 levels on workplace payinequality

Change inpay ratio

Change inworker pay(US$)

Change inmanagerialpay (US$)

Percent unionized −.036 +15.41 +12.79Percent female managers +.049 −.10 +19.85Percent minority workers −.027 +4.45 −2.59

ization levels. I weight calculations by cell frequency, ensuring thatnted in the figures.

in their weekly pay. Worker pay would remain unaf-fected, while the managerial to worker pay ratio wouldbe .05 points higher than where it now stands. If the per-centage of minority workers had not grown during the1980s and 1990s, worker pay would be US$ 4.50 higherwhile managerial pay would be nearly US$ 3 lower.The combined impact of higher worker pay and lowermanagerial pay would have forced the pay ratio down.03 points.

By contrast, the union variable increases both man-agerial and worker pay substantially. Had union levelsnot fallen since 1983, workers would enjoy a weeklyUS$ 15 pay increase while managers would see theirpay rise by US$ 13. Without the post-1983 union decline,the managerial to worker pay ratio would be nearly .04points lower.

Yet high union presence does not raise everyone’s pay.The models in Table 5 measure the impact of unioniza-tion on the pay of upper-level managers. Model 1 ofTable 5 investigates the effect of unions on the ratio of90th percentile managerial to median worker pay. Model2 just examines how unionization levels effect 90thpercentile managerial compensation. Given the instabil-ity of 90th percentile measures with such small countsfor some of the managerial cells, I have dropped thebottom decile of the cell size distribution. Also, prior

to 1989, the CPS’s restrictive topcoding prevents mefrom accurately capturing top-level executive pay. Asa result, both models in Table 5 only include the years1989–2000.

J. Rosenfeld / Research in Social Stratificati

Table 5Unionization effects on top-earner managers

90/50 pay ratio 90th percentilemanager pay

Proportion unionized −1.595* (.685) −138.445 (291.777)

Worker variables% Female .323 (.630) −232.697 (268.062)% Minority .456 (.659) −262.444 (280.559)HS degree −2.079* (.823) −145.903 (350.181)College −2.666** (.913) −26.717 (388.657)Age −.312 (.307) −57.445 (130.633)Age2 .004 (.004) .652 (1.721)

Manager variables% Female −1.390*** (.278) −625.909*** (118.322)% Minority .377 (.419) −31.839 (178.266)Some college −.077 (.399) −134.479 (169.884)College degree .909** (.318) 474.740*** (135.398)Age .283 (.181) 126.560 (77.197)Age2 −.003 (.002) .652 (1.721)

Year fixed-effects Yes YesIndustry/region

fixed-effectsYes Yes

Constant −7.602*** (6.091) 558.368 (2592.876)N 1133 1133Number of

parameters110 110

R2 .68 .79

Note: Standard errors are in parentheses. Model 1 dependent vari-able: a ratio of the 90th percentile managerial pay to median weeklyworker pay. Model 2 dependent variable: 90th percentile managerialpay. Wages in constant US$ 2000. Models weighted by the number ofmanagers in each cell.

* p < .05.** p < .01.

*** p < .001.

Table 6Sensitivity analysis

Model N

(1) Main model (Model 3 from Table 2) 1872(2) Industry and year fixed-effects 1872(3) Industry-region fixed-effects, year linear 1872(4) Region-year and industry fixed-effects 1872(5) Industry-year and region fixed-effects 1872(6) OLS with robust clustered errors by industry-region 1872(7) Firm size effects model 1248(8) PCSE with panel specific AR-1 autocorrelation structure 1872(9) Main model with lagged dependent variable 1768(10) Main model minus panels exhibiting autocorrelation 1530

a p < 10.* p < .05.

** p < .01.*** p < .001.

on and Mobility 24 (2006) 223–238 233

As shown in Model 1, unionization operates to narrowthe gap between median worker and top-level managerialpay. The relationship is significant at the .05 level. Unlikeearlier models, many of the other covariates fail to affectthe ratio, suggesting that top-level managerial pay mayoperate according to a different dynamic than medianmanagerial compensation. This suspicion is confirmedin Model 2, where the relationship between unionizationand the 90th percentile managerial pay measure is non-significant, although negatively signed, meaning that thesignificant negative union coefficient in Model 1 is dueto the union wage premium. Thus, I find no evidence ofa negative relationship between upper-managerial com-pensation and managerial pay, although as discussedpreviously, sample size limitations and topcoding pre-vent me from fully investigating the issue. Alternatively,the measures used in this analysis may be too crude topinpoint the pay determinants of the elite upper strataof the managerial ranks. Unlike the bulk of the occu-pational distribution, top-level executives have enjoyeddramatic wage increases during the prior decades. Itmay indeed be that the determinants of top-level exec-utive pay differ substantially from other occupationalstrata.

Finally, to ensure the findings are robust, Table 6below presents the results of various alternative modelspecifications for my outcomes of interest. Model 1replicates Model 3 from Table 2, the main model ofthe analysis. Models 2–5 introduce different combina-tions of fixed-effects, ranging from separate industryand year effects (Model 2) to region-year and indus-

try fixed-effects (Model 4). None of these combinationsresult in a non-significant manager/worker pay ratio orworker pay coefficient. Only in the model with industry-year and region effects (Model 5) does the managerial

Inequality Worker pay Manager pay

−.610*** 392.541*** 313.371***

−.314*** 319.279*** 326.246***

−.694*** 418.994*** 330.970***

−.514*** 247.204*** 85.562*

−.506*** 172.014*** −54.981−.610** 392.541*** 313.371**

−.395a 211.062*** 177.911a

−.601*** 289.071*** 273.002***

−.482** 222.499*** 292.917***

−.461** 371.138*** 363.760***

atificati

234 J. Rosenfeld / Research in Social Str

pay coefficient change appreciably (it is no longer sig-nificant). However, this is such a stringent (and nearlysaturated) model – with nearly 500 covariates – that theseresults should be taken cautiously (more on Model 5below).

Model 6 of Table 6 displays the results of a truncatedtime-series panel with a firm size measure. Net ofother contributing factors, larger firms offer theiremployees higher wages, although the effect seems tobe diminishing over time (Hollister, 2004). If averagefirm size within particular industry-regions is system-atically related to union presence in the workplace, theexclusion of a firm size measure could bias the effectof unionization on worker and managerial wages. TheCPS MORG data do not contain firm size information;however, beginning in 1989, the March CPS surveyincludes an item asking about employer size. Thediminished effects of this model should be takencautiously: while the firm size coefficient (not shown;available upon request) is significant and positivelyrelated to the managerial/worker pay ratio, runningthe same model minus the firm size variable on the1989–2000 series reveals a diminished union effect onworker pay, indicating that the truncated series accountsfor much of the Model 6 results.6

Given that autocorrelation is often a problem in time-series analyses, Models 8–10 of Table 6 attempt to dealwith the issue in various ways. Despite adding a laggedversion of the dependent variable (Model 9), specify-ing an AR-1 autocorrelation structure (Model 8) andremoving those panels exhibiting unit roots (Model 10),the results remain significant and signed in the expecteddirections.

The results presented in Tables 1–5 differ slightlyfrom those of DiNardo et al. (1997, pp. 20–21). In theirindividual-level analysis of CPS data, many model speci-fications do indicate a significant and positive associationbetween managerial pay and unionization levels. How-ever, their most stringent tests indicate no statisticallysignificant relationship between unionization and man-ager pay (although of the 12 models they estimate, 10are positively signed). Why might their results differfrom my own? Their analysis focuses on wages of allmanagers, and thus does not separate out the effect ofunions on pay for mid-level and upper-level managers.

As discussed previously, determinants of pay for upper-level managers seem to operate according a differentdynamic.

6 A result that, given union’s weakened state in the 1990s comparedto the 1980s, should come as no real surprise.

on and Mobility 24 (2006) 223–238

Also, their most stringent models include a unionwage premium control (not relevant for this study) aswell as a control for an industry wage premium fornon-union workers. While this study controls for timeinvariant industry-region effects, if industry wage differ-entials have changed over time and are correlated withtrends in industry-level unionization rates, the union-ization coefficient may capture some of the effects ofchanging industry wage differentials. Model 5 of Table 6includes a separate industry-year effect and indeed thepositive union-managerial pay relationship disappears.However, this model fails to control for industry-regioneffects and thus fails to account for the importantintra-industry, inter-regional differences in unioneffects.

More importantly, even if the positive effect of unionson median managerial pay that I find partly proxiesfor higher rents (although the models employed in thisanalysis go far in addressing this concern), the over-all story remains the same: unions operate to narrowworkplace wage dispersion. Indeed, if the slight posi-tive union-managerial pay finding works in part throughindustry wage differentials, the negative union effect onworker to manager pay is likely to be slightly underesti-mated.

5. Discussion

While previous research tends to focus on the impactof falling union membership levels on otherwise compa-rable workers (see Kuhn, 1998 for an overview), or oncross-national pay comparisons (DiNardo & Lemieux,1997), this project specifically targets the role of uniondecline on pay inequality across occupations. The find-ings presented above corroborate and expand on theexisting literature in various ways. At the most basiclevel, I demonstrate that income inequality – capturedin this study by a ratio of median managerial to medianworker pay – has increased over time, a finding sug-gested by a host of researchers (see Morris & Western,1999 for a review). In 1983, the ratio of median man-agerial to median worker weekly pay stood at 1.77,by the beginning of the 21st Century the gap hadgrown to 1.90. As income inequality grew – as the gapbetween median managerial pay and median worker paywidened – private sector unionization levels fell drasti-cally. Union membership among private sector, full-timeworkers dropped ten percentage points in an 18-year

period.

Rising inequality and union decline are intimatelyrelated. The significant negative relationship betweenunionization and pay inequality is robust to controls

atificati

ffice

llatirswfrwstciwnmbie

isminfati

togttg(sfTiHdm

J. Rosenfeld / Research in Social Str

or a host of other factors, including various work-orce demographic measures, unmeasured character-stics at the industry-region level and unmeasuredharacteristics captured by specifying a year fixed-ffect.

Simply showing that unionization negatively corre-ates with cross-occupational wage inequality revealsittle about how union membership affects manager paynd worker pay separately. The data analysis showshat unionization increases managerial pay, but reducesnequality by raising worker pay by a larger margin. Thisather unexpected finding lends indirect support to long-tanding union literature demonstrating a broad unionage premium (see Kuhn, 1998 for a review). However,

ew other investigations have revealed how unions canaise wages not only for otherwise comparable workersithin specific industries and regions, but also for certain

upervisory positions. Indeed, the effects of unioniza-ion on managerial pay rival the effects of demographichange (see Table 4). It may be that managers in union-zed firms set their own pay scales with reference to theirorkers’ pay, which will be, on average, higher than inon-unionized firms. Or, the models used in this analysisay not have completely controlled for the association

etween highly unionized firms and rents, since histor-cally unions have clustered in monopoly sectors of theconomy.

Still, despite the positive association between union-zation and median managerial compensation, unionshrink the earnings gap between workers and theiranagers. This finding lends evidence to an under-

nvestigated impact of an organized worker presence:amely, the ways in which unions affect wages for dif-erent occupations. While mid-level managers may seemodest wage boost as their firm unionizes, the rela-

ive gains achieved by workers will more than offset thisncrease.

Unions’ ability to narrow overall wage dispersion inhe workplace lends further evidence to the theory thatrganized labor concerns itself with more than just wageains for its own members. The small positive associa-ion between mid-level managerial pay and unions but-resses this claim: if unions simply engaged in a zero-sumame to capture all resources available for compensationas the lower rents hypothesis suggests), then unionshould lower managerial pay, since any wage increaseor workers must be offset by a wage cut for managers.he fact that unions raise managerial pay while narrow-

ng wage dispersion points to the pay norm hypothesis.owever, it may be that the relative weakness of unionsuring the period under investigation actually allows foranagers in highly unionized locales to capture some

on and Mobility 24 (2006) 223–238 235

of the available rents, and that an analysis into a priorera of relative union dominance would reveal a nega-tive effect on managerial pay. Future research on thematter that focuses on periods when unions were espe-cially dominant could help detangle the pathways (if any)through which unions affected managerial compensa-tion.

Finally, anecdotal evidence discussed earlier in thepaper suggests that unions may act as an institutionalbrake on CEO compensation. The CPS data used in myanalyses do not bear this suggestion out. However, thedata do not explicitly reject this hypothesis either: dueto rather restrictive topcoding, CPS data do not effec-tively capture wage information on top-level executives.Therefore, I am simply unable to confirm whether highlevels of unionization checks CEO and top-level execu-tive compensation. As discussed above, the CPS data doreveal a positive relationship between unionization andlower-level managerial compensation at the industry-region level, but this relationship disappears when Iexamine higher-paid executives. In order to effectivelytest whether unions do indeed push down CEO and top-level executive pay, one would need to link firm-leveldata on executive compensation (readily available formajor firms from a variety of sources, including Forbesand Business Week), with firm-level unionization data.Unfortunately, such firm-level union data are unavail-able, and as this analysis demonstrates, once one aggre-gates up to the industry level, information on top earnersis lost.

Future research on the subject should also incorporatesome measure of union strength beyond sheer mem-bership counts. As prior research shows (Card, 1998;Mitchell, 1994), not only have unions suffered signifi-cant membership losses over the past decades, but evenin those firms and industries that remain highly orga-nized, the general effectiveness of unions seems to haveslackened. Therefore, it is crucial to investigate not onlyhow a decline in the sheer numbers of organized employ-ees affects worker and managerial pay, but also the roleof declining union power. One promising avenue forfuture research may lie in tracking the impact of orga-nized labor on occupational pay for those (still quitefew) unions experiencing actual growth – such as SEIU– and comparing the results to unions hemorrhagingmembers in the last few decades—such as the UAW.It may well prove that certain growing service-sectorunions have greater influence in setting workplace pay

norms than older, industrial labor organizations. Giventhe still quite low union membership rates in service-sector industries, such an analysis will have to wait forsome time.

atificati

236 J. Rosenfeld / Research in Social Str

Finally, as DiNardo et al. (2000) discuss, future anal-yses of unions’ effect on managerial compensation musttry to specifically sort out the role played by higherrents in the correlation between unionization and man-agerial pay. It could be (although the stringent modelspecifications in my investigation go a long way towardeliminating the possibility) that the higher pay for man-agers found in unionized firms simply proxies for higherrents (also associated with higher levels of unionization).While important, even if some of the managerial payboost I attribute to unionization really reflects higherrents, the overall finding that unions act to narrow wagedispersion among workers and managers remains unaf-fected.

6. Conclusion

As the collective bargaining voice for wage labor-ers, unions have traditionally worked to boost averageworker wages—a finding corroborated by this inves-tigation. Moreover, solid evidence suggests that firmsmay raise wages for comparatively similar but non-unionized workers as management acts to stem the threatof further unionization. This paper demonstrates that theunion wage boost spills over into non-unionized posi-tions including mid-level supervisory jobs, effectivelyincreasing pay for a large segment of the workforce.However, in comparison to the effect of union presenceon worker pay, the managerial wage boost remains rel-atively modest. Union presence within the workplaceoperates to narrow the wage distribution, providing evi-dence of the ability of unions to influence overall paynorms within a firm.

The steep declines in union representation over thepast few decades have exacerbated earnings inequalityin the United States. Declining union presence loosensup the pay scale by granting management the decisiveupper hand in bargaining wages and leaving wagework-ers at the mercy of shifting market forces. The dropin wages attributable to declining union presence hascontributed to wage stagnation among workers and mid-level management alike: both segments of the labor forcestand to gain with strong union presence and both havelost ground during this period of unprecedented uniondecline.

A changing workforce demographic composition,continuing deindustrialization and increasing globaliza-tion all play a role in the growing earnings inequality over

the past few decades. However, as this paper demon-strates, research into the recent rise in pay inequalitymust take account of dramatic institutional shifts withinthe labor market. The continuing decline of organized

on and Mobility 24 (2006) 223–238

labor has exacerbated the widening gap between workerpay and managerial pay. Without an abrupt turnaround inunion membership, this gap will only continue to grow.

Acknowledgements

I thank Scott Lynch and Meredith Kleykamp forhelpful assistance on earlier versions of this paper.Very special thanks to Bruce Western for crucialmethodological and theoretical guidance throughoutthis project, and to the two anonymous RSSM reviewersfor their suggestions.

Appendix A

CPS industry groupings (CPS industry recodes pro-vided in parentheses):

1. Agriculture service, other agriculture, forestry andfisheries (01–02, 46).

2. Mining (03).3. Construction (04).4. Lumber, wood, furniture, fixtures, stone, clay, glass

and concrete (05–07).5. Primary and fabricated metals, metal industries not

specified (08–10).6. Machinery, including electric, professional and pho-

tographic equipment, watches (11, 12, 16).7. Auto vehicles, aircraft, parts and other transporta-

tion equipment (13–15).8. Toys, amusement and sporting goods, and miscella-

neous manufacturing (17–18).9. Food, tobacco and kindred products (19, 20).

10. Textiles, apparel and leather products (21, 22, 28).11. Paper, printing, publishing and allied industries (23,

24).12. Chemicals, petroleum and coal products, rubber,

miscellaneous plastic and allied products (25–27).13. Transportation (29).14. Communications (30).15. Utilities and sanitary services (31).16. Wholesale trade (32).17. Retail trade (33).18. Banking and other finance (34).19. Insurance and real estate (35).20. Private household and personal services (36, 39).21. Business and other professional services (37, 45).22. Repair services (38).

23. Entertainment and recreation services (40).24. Hospitals and health services (41, 42).25. Education services (43).26. Social services (44).

atificati

A

1

2

3

4

J. Rosenfeld / Research in Social Str

ppendix B

Regional groupings by state:

. Northeast/mid-Atlantic region- Maine- New Hampshire- Vermont- Massachusetts- Rhode Island- Connecticut- New York- New Jersey- Pennsylvania- Delaware- Maryland- Washington, DC

. South- Virginia- West Virginia- North Carolina- South Carolina- Georgia- Florida- Kentucky- Tennessee- Alabama- Mississippi- Arkansas- Louisiana

. Midwest- Ohio- Indiana- Illinois- Iowa- Michigan- Missouri- Wisconsin- Minnesota- North Dakota- South Dakota- Nebraska- Kansas

. West- Oklahoma- Texas- New Mexico

- Arizona- Montana- Idaho- Nevada

on and Mobility 24 (2006) 223–238 237

- Wyoming- Colorado- Washington- Oregon- California- Alaska- Hawaii- Utah

References

Blanchflower, D., Bryson, A. (2003). What effect do unions have onwages now and would ‘what do unions do?’ Be surprised? NBERWorking Paper 9973. Cambridge, MA: National Bureau of Eco-nomic Research.

Card, D. (1996). The effects of unions on the structure of wages: Alongitudinal analysis. Econometrica, 64, 957–979.

Card, D. (1998). Falling union membership and rising wage inequality:What’s the connection? NBER Working Paper 6520. Cambridge,MA: National Bureau of Economic Research.

Clawson, D., & Clawson, M. A. (1999). What has happened to the USlabor movement? Annual Review of Sociology, 25, 95–119.

Corneo, G., & Lucifora, C. (1997). Wage formation under union threateffects: Theory and empirical evidence. Labour Economics, 4,265–292.

DeAngelo, H., & DeAngelo, L. (1991). Union negotiations and cor-porate policy: A study of labor concessions in the domestic steelindustry during the 1980s. Journal of Financial Economics, 30,3–43.

DiNardo, J., Fortin, N. M., & Lemieux, T. (1996). Labor market institu-tions and the distribution of wages, 1973–1992: A semiparametricapproach. Econometrica, 64, 1001–1044.

DiNardo, J., Hallock, K., Pischke, J. -S. (1997). Unions and ManagerialPay. NBER Working Paper 6318. Cambridge, MA: National Bureauof Economic Research.

DiNardo, J., Hallock, K., Pischke, J. -S. (2000). Unions and the LaborMarket for Managers. CEPR Discussion Paper 2418. London, UK:Centre for Economic Policy Research.

DiNardo, J., & Lemieux, T. (1997). Diverging male wage inequality inthe United States and Canada, 1981–1988: Do institutions explainthe difference? Industrial & Labor Relations Review, 50, 629–651.

Freeman, R. B. (1993). How much has de-unionization contributed tothe rise in male earnings inequality? In S. Danziger, & P. Gottschalk(Eds.), Uneven tides: Rising inequality in America (pp. 133–163).Russell Sage Foundation.

Freeman, R. B., & Medoff, J. (1984). What do unions do? Basic Books.Goldin, C., Katz, L. F. (1999). The returns to skill across the twentieth

century United States. NBER Working Paper 7126. Cambridge,MA: National Bureau of Economic Research.

Grusky, D. B., & Sorenson, J. B. (1998). Can class analysis be sal-vaged? American Journal of Sociology, 103, 1187–1234.

Hirsch, B. T. (2004). Reconsidering union wage effects: Surveying newevidence on an old topic. Journal of Labor Research, 25, 233–266.

Hollister, M. N. (2004). Does firm size matter anymore? The new econ-

omy and firm size wage effects. American Sociological Review, 69,659–676.

Kalleberg, A., Wallace, M., & Althauser, R. (1981). Economic segmen-tation, worker power, and income inequality. American Journal ofSociology, 87, 651–683.

atificati

the USA: A comparison of the 1960s and 1990s. Socio-Economic

238 J. Rosenfeld / Research in Social Str

Kuhn, P. (1998). Unions and the economy: What we know; What weshould know. Canadian Journal of Economics, 31, 1033–1056.

Kuhn, P., & Sweetman, A. (1998). Wage loss following displacement:The role of union coverage. Industrial and Labor Relations Review,51, 384–399.

Leicht, K. T. (1989). On the estimation of union threat effects. Ameri-can Sociological Review, 54, 1035–1047.

Leicht, K. T., Wallace, M., & Grant, D. S., II. (1993). Union presence,class, and individual earnings inequality. Work and Occupations,20, 429–451.

Lewis, H. G. (1986). Union relative wage effects: A survey. Universityof Chicago Press.

Martin, S., & Rence, C. (1984). Vertical spillovers, market concentra-tion, union coverage, and wages. Journal of Labor Research, 5,177–189.

Milkman, R. (1987). American women and industrial unionism duringWorld War II. In M. R. Higonnet, J. Jenson, S. Michel, & M. C.Weitz (Eds.), Behind the lines: Gender and the two world wars (pp.168–181). Yale University Press.

Mitchell, D. J. B. (1994). A decade of concession bargaining. In C.Kerr, & P. D. Staudohar (Eds.), Labor economics and industrialrelations: Markets and institutions (pp. 435–474). Harvard Uni-versity Press.

Morris, M., & Western, B. (1999). Inequality in earnings at the close ofthe twentieth century. Annual Review of Sociology, 25, 623–657.

Neumark, D., & Wachter, M. L. (1995). Union effects on nonunionwages: Evidence from panel data on industries and cities. Industrialand Labor Relations Review, 49, 20–38.

Perrone, L. (1983). Positional power and propensity to strike. Politicsand Society, 12, 231–261.

on and Mobility 24 (2006) 223–238

Perrone, L. (1984). Positional power, strikes, and wages. AmericanSociological Review, 49, 412–421.

Rees, A. (1989). The economics of trade unions. University of ChicagoPress.

Robinson, C. (1989). The joint determination of union status and unionwage effects. Journal of Political Economy, 97, 639–667.

Singh, P., & Agarwal, N. C. (2002). Union presence and executivecompensation: An exploratory study. Journal of Labor Research,23, 631–646.

Sorenson, A. B. (1996). The structural basis of social inequality. Amer-ican Journal of Sociology, 101, 1333–1365.

Stephan-Norris, J., & Zeitlin, M. (2003). Left out: Reds and America’sindustrial unions. Cambridge University Press.

Swenson, P. (1989). Fair shares: Unions, pay, and politics in Swedenand West Germany. Cornell University Press.

Wallace, M., Griffin, L. J., & Rubin, B. A. (1989). The positional powerof American labor, 1963–1977. American Sociological Review, 54,197–214.

Western, B., & Healy, K. (1999). Explaining the OECD wage slow-down: Recession or labour decline? European Sociological Review,15, 233–249.

Wong, E. (2003). Under fire for perks, chief quits American Airlines.The New York Times [April 25].

Wright, E. O., & Dwyer, R. E. (2003). The patterns of job expansions in

Review, 1, 289–325.Zeitlin, M. L., & Weyher, F. (2001). ‘Black and white, unite and fight’:

Interracial working-class solidarity and racial employment equal-ity. American Journal of Sociology, 107, 430–467.