unions and interindustry wage differentials

Unions and Interindustry Wage Di�erentials

JORGE SABA ARBACHEUniversidade de Brasila, Brasilia, Brazil

and

FRANCISCO GALRAO CARNEIRO *

Universidade Catolica de Brasila, Brasilia, Brazil

Summary. Ð We investigate the importance of trade unions in collective bargaining in thecontext of a developing country manufacturing labor market. The methodology weadopt to estimate wage di�erentials follows the method proposed by Haisken-DeNew,J. P. and Schmidt, C. M. (1997) Review of Economics and Statistics 79, 516±521, since itimproves on the standard procedure popularized by Krueger, A. B. and Summers, L. H.(1988) Econometrica 56, 193±259. Our ®ndings indicate that wage dispersion is fargreater in the unionized sector of Brazilian manufacturing, in contrast to evidence fromother countries. Ó 1999 Elsevier Science Ltd. All rights reserved.

1. INTRODUCTION

The structure of interindustry wages was thefocus of intense research in the early 1950s andhas received a great deal of attention in recentyears. Some of the stylized facts found by earlierstudies include evidence of large di�erentials inwages across industries, even after controls havebeen implemented (Slichter, 1950), and aremarkable temporal persistence of the interin-dustry wage structure (e.g., Allen, 1995). Manyauthors have tried to argue that these stylizedfacts su�ce to rule out the adequacy of simplecompetitive models of the labor market againstthe superiority of alternative approaches ofnoncompetitive models such as e�ciency wage,rent-sharing or insider-outsider models (e.g.,Holmlund and Zetterberg, 1991, and Blanch-¯ower, Oswald and Sanfey, 1996).

A growing line of research has focusedattention on the evidence of a union-nonunionwage di�erential (e.g., Booth, 1995). Models ofthe aggregate labor market, for example, haveused the evidence of such a di�erential as aproxy for union power. The underlying featureseems to be the existence of economic rents,which can be bargained for. Higher unionwages re¯ect the ability of trade unions to force®rms to give up some of their surplus. In this

sense, the union wage di�erential would bepositively related to union power, as assumedin aggregate labor market models (Layard andNickell, 1986).

In the study of developed country labormarkets, the sources of these rents are usuallyassociated with the size of the ®rms, industrya�liation, and productive performance (e.g.,Blanch¯ower, Oswald and Sanfey, 1996). Fordeveloping countries, however, there aredi�erent accounts on the source of these rents.Teal (1996), for example, suggests that thepublic sector would be the primary source ofrents in developing country labor marketsbecause of their characteristic of strong insti-tutional in¯uence in wage determination. Thepublic sector would be in general determiningan overall wage policy as a ¯oor to wagenegotiations and also establishing a distinctionwithin the own public sector between produc-tive and unproductive activities (Gelb, Knightand Sabot, 1991).

World Development Vol. 27, No. 10, pp. 1875±1883, 1999Ó 1999 Elsevier Science Ltd. All rights reserved

Printed in Great Britain0305-750X/99/$ - see front matter

PII: S0305-750X(99)00090-Xwww.elsevier.com/locate/worlddev

* We are grateful to two anonymous referees, Ricardo

Paes e Barros, Andy Dickerson, Jo~ao R. Faria, Marcelo

Neri, and Peter Sanfey for very helpful comments and

suggestions. The usual disclaimer applies. Final revision

accepted: 28 March 1999.

1875

There is also some evidence that liberalizationof collective bargaining in the context of adeveloping country facing an intermediatecentralized bargaining structure1 may be ratherconducive to rent sharing (Carneiro, 1998).Powerful wage bargainers would face littleresistance to convert positive demand shocksinto wage gains and also to appropriate tothemselves productivity improvements in detri-ment to the rest of society. In this sense, asargued by Booth (1995), the union wage di�er-ential over nonunion alternatives is expected tobe increasing with union power.

In this paper we examine the role of unions inbringing about interindustry wage di�erentials.Previous accounts on the role of unions in theprocess of wage determination in Brazil havefocused on: (a) the growth of insider power andthe decreasing importance of the government incollective bargaining (Carneiro and Henley,1998); (b) the temporal stability of the interin-dustry wage structure (Gatica, Mizala andRomaguera, 1995); and (c) the role of humancapital factors in explaining wage di�erentials(Arbache, 1997). The assessment of how tradeunions contribute to overall wage dispersion inBrazil,however, isyet tobedone.Unlikeprevious®ndings (Gosling and Machin, 1995; Fortin andLemieux, 1997), our results indicate that underthe intermediate centralized bargaining structurepresent in Brazil2 trade unions contribute toincreasing rather than decreasing wage disper-sion within the union sector.

The paper is structured as follows. In the nextsection, we present the theoretical backgroundunderlying our analysis of union wage di�eren-tials. In Section 3 we discuss the data used inthis paper. In Section 4 we analyze severalfeatures of the data, exploiting the relevance ofunion power to explain the existence of a unionwage di�erential. Section 5 concludes the paperand discusses some of the possible implicationsof the results.

2. THEORETICAL BACKGROUND

Before considering the empirical evidence, wewill ®rst discuss how unions can have an impactupon wage determination and the consequencesof di�erent bargaining structures in shapingdi�erent outcomes in the labor market.

A recent development in the economics of thelabor market was the appearance during the1980s of a wide range of studies on the behaviorof trade unions and their impact on the process

of wage and employment determination. In thespeci®c case of wage determination, it has beennow widely suggested that wage moderation isusually associated with a more coordinatedbargaining structure, or what has been termedcorporatist arrangements. On the other extreme,however, explosive wage demands have beenusually associated both with a lack of coordi-nation and with synchronization in wagebargaining.

The discussion on the role of institutions andwage bargaining structures in determining wagemoderation cannot be dissociated from thedebate on labor market ¯exibility. In thecompetitive model of the labor market, forexample, the balance of supply and demand inthe whole market dictates wages. Hicks's Theoryof Wages (1932) was the starting point for thisargument, but even Hicks himself becameskeptical about the overall applicability of thecompetitive model, as he made clear in thepreface to Hicks (1963). There has been muchdebate whether wages can be considered as ®xedby noncompetitive pressures. A common argu-ment often used is that in most cases wages aredetermined by collective bargaining and alsothat other noncompetitive factors might beimportant (e.g., e�ciency wage and rent-sharingtheories).

The seminal contribution on the debateconcerning the ability of organized groups toa�ect economic performance is attributed toOlson (1965, 1982). His view is that the ®nale�ect of the action of interest groups in a societyis mostly negative. Olson is particularlyconcerned with the fact that interest groups tendto act in detriment to the rest of the society byappropriating any improvement in e�ciencythat their collective action might have gener-ated. The accumulation of these groups mayultimately increase the complexity of regulation,the role of the government, reducing allocativee�ciency and thus the rate of economic growth.His preferred solution, therefore, seems to be areduction in the capacity of economic intereststo organize at all.

Crouch (1985) argues that Olson's perceptionleads to a paradox. In Olson's view, onceeconomic interests have become organized, theywill be less likely to act in line with the idea ofallocative e�ciency. But as the level of organi-zation increases these special-interest groupsmay broaden their interests and become morepoliticized, aggregated and centralized. Theseencompassing organizations may then includeconcerns about economic growth among other

1876 WORLD DEVELOPMENT

interests that they might also have (e.g.,regarding the politics of income distribution).Considering the case of trade unions, Crouchsuggests that the more centrally united isorganized labor, the more are its actionscompatible with the stability of the marketeconomy (see also Bruno and Sachs, 1985).

By investigating if, among the several char-acteristics that make up corporatism, central-ization was empirically relevant, Calmfors andDri�ll (1988) were able to suggest that theclaimed monotonic relationship betweencentralization and economic performance wasnot always true. Their results implied the exis-tence of a hump-shaped relationship betweenunemployment and bargaining centralization.

In the polar case of decentralization, negoti-ations between workers and ®rms are the ulti-mate determinants of wages; the fact that agentsare too small in this context to interfere in theoperation of the market means that wagerestraint holds because competitive pressuresput a cap on price and wage increases. At theother extreme of centralized, economy-widebargaining, the idea is that corporatist arrange-ments may serve to increase the awareness ofwage setters to the macroeconomic conse-quences of their actions. In periods of adverseshocks, corporatist institutions may thuscontribute to reduce uncertainty regarding thedeleterious consequences of possible distribu-tional con¯ict and facilitate a relatively painlessadjustment process.

In the case of the intermediate group, wherebargaining takes place mostly at the industrylevel, the explanation for the poorer perfor-mance relative to the two polar cases is usuallythe existence of poorly coordinated monopo-listic power in the labor market that constrainsthe successful operation of either competitivemarket forces or corporatist coordination. Inthis context, trade unions' bargaining powermay be quite strong since rapid labor turnoveris restricted by high training costs and or hiringand ®ring costs. Insider bargaining power isalso high, meaning that workers perceive thatby pressing for higher wages the ®rm's productprice may increase without however a�ectingthe aggregate price level and therefore their realtake home pay. As bargaining is at the industrylevel, an individual ®rm will not achieve acompetitive advantage over the others and,assuming that industry demand is relativelyinelastic, all the ®rms believe that the overallemployment consequences will be insigni®cant.Overall, therefore, the situation of intermediate

collective bargaining with strong trade unionsmay be quite conducive to poor employmentand in¯ation performance, with rent sharingbeing a pervasive characteristic.

Freeman (1988) and Rowthorn (1992) showthat wage dispersion also tends to be lower incorporatist economies as opposed to economieswith a decentralized labor market. In the caseof an economy with an intermediate centralizedbargaining structure, one should thus expectgreater dispersion relatively to the polar cases.As the union wage di�erential tends to increasewith union power, the intermediate centralizedbargaining structure seems to favor the exis-tence of a rather large wage dispersion. In whatfollows, we present some evidence on the caseof Brazil with the aim of assessing the validityof this argument.

3. DATA DESCRIPTION

The micro-data used in our investigation arefrom the National Household Surveys(Pesquisa Nacional por Amostragem Domiciliar± PNAD) of 1992 and 1995, which areconducted by the Brazilian Institute of Geog-raphy and Statistics (IBGE). The sample we useis ®ltered to include economically active indi-viduals, nonemployers, aged between 18 and 65years, who work in their main occupation, withformal labor contract, and are a�liated to anyof the 22 industries that compose the manu-facturing sector.

In Table 1 we present the means for anumber of variables in both subsamples.Unionized workers seem to accumulate morehuman capital characteristics as illustrated bytheir means regarding education, tenure andexperience. The subsample of nonunionizedworkers is made up of relatively more femalesand nonwhites, and who are paid less nonwagebene®ts. Nonunion workers are also likely to befound in the poorest regions, occupying lowand medium-skill occupations, and are subjectto more overtime work.

Table 2 presents data on union density inthe manufacturing sector. In general, uniondensity varies from industry to industry. Trans-port material, mechanic, metallurgic, elec-tronics, paperÐcapital-intensive industriesÐpresent union densities above the weightedmean. On the other hand, traditionally labor-intensive industries such as apparel, furniture,nonmetallic, and food present union densitiesbelow the weighted mean.

UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS 1877

According to Lipietz (1997), there are somestylized facts indicating a relationship between®rm size and union density. In large plants,where Fordist and Taylorist managerialprocedures are usually applied, workers tend tobe more organized and unionism tends to¯ourish. In order to investigate this proposi-tion, we have calculated the Pearson correla-tion coe�cient (two-tailed) between the uniondensity and industry ®rm size mean.3 The esti-mated coe�cient for 1992 is 0.6244 (p� 0.002),thus suggesting that union density is related to®rm size.

So far, therefore, the two subsamples provideevidence of some distinct characteristics forunionized and nonunionized workers. Aninteresting question at this point is whether

wage di�erences between union and nonunionworkers can be attributable not to unions, butto di�erences in observable productive traits,occupation, and regional characteristics. In thiscase, unions would not a�ect wage determina-tion and would not be responsible for theobservable wage di�erences. In order to assessthis point, the next section investigates wagedi�erentials for both subsamples.

4. UNIONS AND WAGE DIFFERENTIALS

The methodology we adopt to estimate wagedi�erentials follows the method proposed byHaisken-DeNew and Schmidt (1997), thatimproves the standard procedure popularized

Table 1. Means for selected variables for union and nonunion workers

Variables 1992 1995

Unions Nonunions Unions Nonunions

Education (years) 7.8 7.25 8.41 7.66Experience (years) 19.88 18.36 19.5 18.42Experience square 520.41 468.54 499.27 471.16Tenure (years) 6.34 4.05 6.61 4.11Gender (female� 1) 0.2281 0.281 0.2191 0.2721Married (married� 1) 0.7164 0.6025 0.7069 0.6077Head of family (� 1) 0.6666 0.564 0.6665 0.5706Race (white� 1) 0.6434 0.6034 0.6672 0.6046Metropilitan area (� 1) 0.3343 0.328 0.3323 0.3265Ovetime worked (48 h/week� 1) 0.2523 0.3123 0.2437 0.3154

OccupationManager 0.0787 0.0701 0.0828 0.0766Professionals 0.0476 0.0336 0.0650 0.0317Clerical 0.0898 0.1010 0.0833 0.0894Sales 0.0232 0.0467 0.0236 0.0543Skilled a 0.0708 0.0395 0.0701 0.0404Semi-skilled a 0.5591 0.5363 0.5592 0.5349Unskilled a (� 1) 0.1310 0.1728 0.1161 0.1727

RegionSouth 0.3347 0.2551 0.3079 0.2483Southeast 0.4626 0.4948 0.4776 0.4957North 0.0348 0.034 0.0295 0.0411Centrewest 0.0219 0.0502 0.0219 0.0596Northeast (� 1) 0.1460 0.1658 0.1631 0.1553

Non-wage bene®tsHousing 0.0330 0.0525 0.0260 0.0388Meals 0.5110 0.3606 0.6173 0.4552Transportation 0.5544 0.4765 0.6032 0.5187Education/training 0.0597 0.0221 0.0772 0.0308Health 0.4128 0.2523 0.5078 0.2843N 4.0550 6.7300 4.1080 6.8110

a Skilled� craftsmen, foremen and kindred workers; Semi-Skilled� operatives and kindred workers; andUnskilled� service workers.


by Krueger and Summers (1988).4 The wageequations are estimated in the following form:

ln wij � a� bXi � uZj � eij; �1�where w is the natural logarithm of the hourlyreal wage of worker i in industry j, X is thevector of personal characteristics, occupationsand regions, Z is the vector of industrydummies which includes all industries, a is theintercept term, e is a random disturbance termre¯ecting unobserved characteristics and theinherent randomness of earnings statistics, andb and u are the vectors of parameters to beestimated. Since in this model the crossproductmatrix of the regressors is not of full rank, alinear restriction is imposed on the us asfollows,X

ujnj � 0; �2�where n is the employment share in industry j.The reported coe�cients are interpreted as thedi�erence between the mean log wage for aworker in industry j and the employment-shareweighted mean log wage for workers in theentire sample (measured as the employment-share weighted mean of the coe�cients on allindustry dummy variables). The formulationgiven by Eqns. (1) and (2) provides botheconomically sensible coe�cients and their

correct standard error in a single regressionstep.

The standard deviation of wage di�erentialsis computed as

SD�u� ��n0� �H�uj��uj ÿ n0 �D�V �uj��

q: �3�

SD�u� gives the weighted and adjusted stan-dard deviation of coe�cients, �H (.) transforms acolumn vector into a diagonal matrix whosediagonal is given by the column vector itself, �Ddenotes the column vector formed by thediagonal elements of a matrix, and V is thevariance-covariance matrix.

(a) Industry e�ects and union e�ects

Table 3 presents coe�cient estimates ofinterindustry wage di�erentials and the calcu-lated standard deviation of wage spread. Aftercontrols for human capital, demographic char-acteristics, occupations, regions and nonwagebene®ts have been applied, we can still observesome wage dispersion for both unionizedand nonunionized workers.5 Capital-intensiveindustries6 tend to pay a wage premium, whiletraditional industries tend to pay lower wagesfor workers with apparently the same observ-able productive attainments and other charac-teristics as those in the capital-intensiveindustries.

Despite the di�erences in the characteristicsof unionized and nonunion workers, the pro®leof wage di�erentials seems to be similar acrossthe groups. The estimated Pearson correlationcoe�cients are 0.7796 (p� 0.000) for 1992, and0.7094 (p� 0.000) for 1995, suggesting that,irrespective of the union status, there is anindustry e�ect in wage determination; i.e., wagepremia are paid to all workers regardless oftheir a�liation to a trade union. In line withPencavel (1991), this result might be viewed asthe case in which union activism provokesspillovers over other sectors of the economy.Nonunion sectors would try to mimic the wagepractices in the unionized sectors by fearingturnover costs, bad morale, and even the threatof unionization (Dickens, 1986).

When we investigated this result further byrunning wage equations based on an unsplitsample we found a signi®cant union e�ect inmanufacturing wage determination. Actually,the union dummy was statistically signi®cantfor both years (0.1070; t-value� 10.12, for1992; and 0.0656, t-value� 6.39 for 1995),suggesting that despite the common structure

Table 2. Union density

Industry 1992 1995

Apparel 33.2 33.6Beverages 34.5 37.7Chemical 38.2 37.4Electronic 45.0 42.1Food 30.3 29.3Furniture 25.4 20.3Leather 39.7 43.6Mechanic 43.8 44.3Metallurgic 44.3 45.9Mineral 44.9 47.8Nonmetallic 28.6 28.7Other 29.9 30.2Paper 47.5 52.1Perfumes 19.5 26.1Pharmaceutical 33.6 37.1Plastic 31.7 31.0Publishing 36.2 34.4Rubber 32.1 46.8Textiles 50.5 55.7Tobacco 63.9 45.7Transport 51.0 51.9Wood 28.6 21.5Weighted mean 37.6 37.6


of union and nonunion wages, unions have anon-negligible role in wage determination. Bytheir anti-logs, these coe�cients show that theunion wage di�erential in Brazil was of roughly11.3% in 1992, and 6.7% in 1995.7 These ®guresare quite close to the international evidencesurveyed by the World Bank (1995), p. 81). Indeveloping countries like Mexico and Malaysia,the union wage union premium is estimated inthe range of 10±20%. Similar ®gures are foundfor developed countries as well, as surveyed byBooth (1995), chapter 6). The union wagepremium in Germany and the United King-dom, for example, was estimated at 10% and5%, respectively, during 1985±87. Theoreticaljusti®cations for this result can be searched inthe monitoring and turnover models of e�-ciency wages (see Shapiro and Stiglitz, 1984,and Stiglitz, 1985, respectively), which do notdistinguish the union status of workers, or innormative e�ciency wage models that predict

the importance of employers' ability-to-pay(Akerlof, 1984).

In order to assess the relevance of tradeunions in a�ecting wage di�erentials' we haveestimated the weighted and adjusted standarddeviation of wage di�erentials.8 The results atthe bottom of Table 3 show that: (i) the wagedispersion coe�cient found for unionizedworkers is greater than that found fornonunionized workers, and (ii) the relativedi�erence between the measures of dispersionincreases over 1992±95. This stresses the pointthat trade unions cause greater dispersion ofwages in the case of Brazil; a result which holdseven under changes in the economic contextwith greater economic openness, lower in¯ationand increased competitiveness after 1994.Variations in the observed union wage gapacross sectors are likely to be related to di�er-ences in the surplus made available by ®rms,and to the ability of trade unions to appropri-

Table 3. Wage di�erentials for union and nonunion workers (standard errors in parentheses) a

1992 1995

Union Nonunion Union Nonunion

Apparel ÿ0.2081 (0.0260) ÿ0.0959 (0.0178) ÿ0.1502 (0.0249) ÿ0.0926 (0.0175)Beverages ÿ0.0462 (0.0536) ÿ0.0514 (0.0381) ÿ0.0067 (0.0480) ÿ0.0026 (0.0372)Chemical 0.3162 (0.0323) 0.1228 (0.0247) 0.1589 (0.0327) 0.0643 (0.0249)Electronic 0.1189 (0.0360) 0.0925 (0.0328) 0.0265 (0.0345) 0.0605 (0.0295)Food ÿ0.1385 (0.0214) ÿ0.1080 (0.0134) ÿ0.1495 (0.0204) ÿ0.0750 (0.0127)Furniture ÿ0.2217 (0.0546) ÿ0.0654 (0.0311) ÿ0.0666 (0.0592) ÿ0.0710 (0.0294)Leather ÿ0.3132 (0.0695) ÿ0.0710 (0.0553) ÿ0.2179 (0.0757) ÿ0.1772 (0.0663)Mechanic 0.0462 (0.0312) 0.0837 (0.0272) 0.0430 (0.0288) 0.0640 (0.0256)Metallurgic 0.0553 (0.0203) 0.1146 (0.0180) 0.0498 (0.0192) 0.0791 (0.0178)Mineral 0.2445 (0.0444) 0.0557 (0.0391) 0.2107 (0.0409) 0.0488 (0.0385)Nonmetallic 0.0120 (0.0384) ÿ0.0073 (0.0238) ÿ0.0886 (0.0357) ÿ0.0222 (0.0225)Other ÿ0.0861 (0.0659) ÿ0.0664 (0.0420) ÿ0.0124 (0.0530) ÿ0.0265 (0.0345)Paper 0.0888 (0.0479) 0.1005 (0.0444) ÿ0.0132 (0.0452) 0.0422 (0.0471)Perfumes 0.0769 (0.1272) 0.0255 (0.0610) 0.1552 (0.1011) ÿ0.0270 (0.0595)Pharmaceutical 0.1612 (0.0831) 0.1109 (0.0576) 0.0379 (0.0819) 0.2154 (0.0623)Plastic ÿ0.0582 (0.0525) 0.0091 (0.0349) ÿ0.1274 (0.0517) 0.0153 (0.0344)Publishing ÿ0.0214 (0.0400) 0.0266 (0.0294) 0.1520 (0.0400) 0.0761 (0.0284)Rubber 0.0458 (0.0845) 0.0101 (0.0568) 0.0921 (0.0628) 0.1020 (0.0587)Textiles ÿ0.1018 (0.0270) ÿ0.0122 (0.0270) ÿ0.1250 (0.0279) 0.0119 (0.0314)Tobacco 0.1215 (0.0814) 0.2701 (0.1063) 0.1293 (0.1057) 0.0817 (0.0965)Transport 0.2074 (0.0293) 0.2561 (0.0291) 0.2374 (0.0255) 0.1555 (0.0262)Wood ÿ0.1210 (0.0510) ÿ0.1324 (0.0317) ÿ0.0943 (0.0542) ÿ0.0706 (0.0281)R2 0.5582 0.4959 0.6605 0.6157F 110.09 142.94 209.36 235.63SD 0.1329 0.0557 0.105 0.0099n 4.055 6.730 4.108 6.811

a Dependent variable is hourly real wage. Independent variables are education, experience, experience square,gender, marital status, head of family, race, metropolitan residence, seven occupational status dummies(unskilled� 1; see Table 1 for the de®nition of occupational status), ®ve nonwage bene®ts dummies, tenure, overtimework, and ®ve regional dummies (Northeast� 1).


ate some of these rents. The varying bargainingpower of di�erent trade unions ampli®es wagedispersion for workers with apparently thesame characteristics.

(b) Further testing on union power

In this section, we exploit further the datausing the Blinder±Oaxaca decompositionmethodology (see Blinder, 1973, and Oaxaca,1973) to distinguish between the portion ofwage di�erences that is due to observablecharacteristics and the portion that is not. As inBooth (1995), we assume that the unexplainedwage gap is due to union power.

De®ne raw wage di�erentials asD � ln �wU ÿ ln �wN , where �w is the average wageof U, unionized workers, and N, nonunionizedworkers. Considering the wage equations esti-mated so far, we compute the following:

DI � �aU ÿ aN � �X�lU � �X U ÿ �X N ��

�X� �X N �lU ÿ lN ��; �4�

DII � �aU ÿ aN � �X�lN � �X U ÿ �X N ��

�X� �X U �lU ÿ lN ��; �5�

where a is the intercept, X the vector of exog-enous variables for each subsample9 and l thevector of estimated coe�cients. In the absenceof a union e�ectÐwhich is given by di�erentpayo�s to productive traitsÐwage di�erentialswould be explained only by di�erences in meanvalues of human capital and other variables.10

Table 4 reports the results of such decom-position analysis. In 1992, roughly 29.6%(calculated as (0.1381 ÿ 0.0387)/0.3357), whilein 1995 approximately 18.8% of raw wagedi�erentials between unionized and nonunion-ized workers are not explained by di�erences inobservable productive endowments, occupa-tion and other variables,11 con®rming ourprevious ®ndings that unions have an impor-tant role in wage determination. More impor-

tantly, however, these results state that, all elseheld constant, when a worker changes his/herunion membership status from nonunion tounion member s/he is liable to perceive a wageincrease.

It should be noted that our results contradictrecent empirical evidence that attribute to tradeunions the capacity to reduce rather thanincreasing wage dispersion within the unionsector (e.g., Stewart, 1995; Gosling and Machin,1995; Fortin and Lemieux, 1997). The ability toreduce wage dispersion seems to be moreapparent in contexts in which labor marketsoperate freer of institutional rigidities (Free-man, 1988; Rowthorn, 1992). In cases wherebargaining takes place at an intermediatecentralized level, there is a considerable di�er-ence in the bargaining power of di�erent unionsand this is believed to favor an increasing,rather than a decreasing, spread of earningswithin the union sector. Our results, therefore,appear compatible with the structure of manu-facturing labor markets in Brazil.

5. CONCLUSIONS

In this paper, we have investigated theimportance of trade unions in collectivebargaining in the context of a developingcountry manufacturing labor market. Our®ndings indicate that, in the case of Brazil,unions contribute to increasing rather thandecreasing the spread of wages within the unionsector. Moreover, for the case of Brazil, achange in membership status for manufactur-ing workers seems to be relevant, as unionizedworkers tend to earn higher wages. Previousstudies on the union-nonunion wage di�erentialin developed countries, particularly the UnitedKingdom and the United States, have foundlower dispersion within the union sector. In linewith the earlier theoretical discussion, we haveargued that the evidence presented here for thecase of Brazil may be related to the structure of

Table 4. Decomposition analysis of wage di�erences a

Year Di�erential Intercept Method Endowments Coe�cients

1992 0.3357 0.1381 I 0.2363 ÿ0.0387II 0.233 ÿ0.0354

1995 0.3575 0.1895 I 0.2902 ÿ0.1222II 0.2935 ÿ0.1255

a Estimations based on means and coe�cients respectively from Tables 1 and 3.


collective bargaining prevailing in Brazilianmanufacturing.

More speci®cally, we have found a positivecorrelation between ®rm size and union densityin Brazil, and also that wage premia are paid toall workers irrespective of their a�liation to atrade union. Further investigation of the datarevealed however that, despite the apparentlycommon structure of union and nonunion

wages, unions play a non-negligible role inbringing about interindustry wage di�erentials.Our estimations show that wage dispersion isgreater for unionized workers and that therelative di�erence between the measures ofdispersion increases over 1992±95. Further-more, our results state that, in Brazil, when aworker becomes a�liated to a trade union s/heis liable to perceive a wage increase.

NOTES

1. According to the literature that has been popular-

ized by Calmfors and Dri�ll (1988), an intermediate

centralized level of collective bargaining occurs when

bargaining takes place mostly at the industry level.

2. A discussion of Brazilian wage setting institutions

can be found in Carneiro (1998) and Carneiro and

Henley (1998).

3. The data on industry ®rm size mean are from

Relat�orio Anual de Informacß~oes Sociais (RAIS),

conducted by the Ministry of Labour. We have not

computed correlation coe�cients for 1995 because the

data were not available for that year.

4. Haisken-DeNew and Schmidt show that the Krue-

ger and Summers procedure overstates substantially the

standard error of coe�cients, and depending on the

choice of industry chosen as reference, the problem can

be even more apparent. As a consequence, it can lead to

spurious inferences regarding individual elements of the

wage di�erentials vector. In addition, the estimated

variance of the renormalized coe�cients is overesti-

mated in their procedure, which a�ects the estimation of

summary measures such as standard deviation of coef-

®cients.

5. It could be argued that a number of the parameters

estimated might di�er signi®cantly by gender, and that is

indeed a possibility. In the case of Brazilian manufac-

turing, however, Arbache (1999) shows that these results

hold for a sample of male workers and also for a sample

of female workers.

6. The following industries are usually considered as

capital-intensive: chemicals, electronic, mechanic, metal-

lurgic, pharmaceutical, transport.

7. The drop in wage di�erentials during 1992±95 might

be due to the increasing openness of the Brazilian

economy and also to a tighter labor marketÐtwo factors

that tend to reduce union bargaining power.

8. Standard deviations are calculated using Eqn. (3).

9. The vector X in Eqns. (4) and (5) includes industry

dummy variables.

10. To avoid outliers having undue in¯uence in the

determination of the pro®le of an average worker in each

sector, the within-sector median was used for all

continuous variables, while the mean was used for all

dummy variables.

11. Unmeasured abilities and other e�ects not captured

by our dataset may also a�ect this result, which can be

thus either over or underestimated.

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