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Unions and Interindustry Wage Differentials
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World Development Vol. 27, No. 10, pp. 1875±1883, 1999 Ó 1999 Elsevier Science Ltd. All rights reserved Printed in Great Britain 0305-750X/99/$ - see front matter
PII: S0305-750X(99)00090-X
Unions and Interindustry Wage Dierentials JORGE SABA ARBACHE Universidade 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 the
context of a developing country manufacturing labor market. The methodology we adopt to estimate wage dierentials follows the method proposed by Haisken-DeNew, J. P. and Schmidt, C. M. (1997) Review of Economics and Statistics 79, 516±521, since it improves 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 far greater in the unionized sector of Brazilian manufacturing, in contrast to evidence from other countries. Ó 1999 Elsevier Science Ltd. All rights reserved.
1. INTRODUCTION The structure of interindustry wages was the focus of intense research in the early 1950s and has received a great deal of attention in recent years. Some of the stylized facts found by earlier studies include evidence of large dierentials in wages across industries, even after controls have been implemented (Slichter, 1950), and a remarkable temporal persistence of the interindustry wage structure (e.g., Allen, 1995). Many authors have tried to argue that these stylized facts suce to rule out the adequacy of simple competitive models of the labor market against the superiority of alternative approaches of noncompetitive models such as eciency 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 focused attention on the evidence of a union-nonunion wage dierential (e.g., Booth, 1995). Models of the aggregate labor market, for example, have used the evidence of such a dierential as a proxy for union power. The underlying feature seems to be the existence of economic rents, which can be bargained for. Higher union wages re¯ect the ability of trade unions to force ®rms to give up some of their surplus. In this
sense, the union wage dierential would be positively related to union power, as assumed in aggregate labor market models (Layard and Nickell, 1986). In the study of developed country labor markets, the sources of these rents are usually associated with the size of the ®rms, industry aliation, and productive performance (e.g., Blanch¯ower, Oswald and Sanfey, 1996). For developing countries, however, there are dierent accounts on the source of these rents. Teal (1996), for example, suggests that the public sector would be the primary source of rents in developing country labor markets because of their characteristic of strong institutional in¯uence in wage determination. The public sector would be in general determining an overall wage policy as a ¯oor to wage negotiations and also establishing a distinction within the own public sector between productive and unproductive activities (Gelb, Knight and Sabot, 1991).
* 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.
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There is also some evidence that liberalization of collective bargaining in the context of a developing country facing an intermediate centralized bargaining structure1 may be rather conducive to rent sharing (Carneiro, 1998). Powerful wage bargainers would face little resistance to convert positive demand shocks into wage gains and also to appropriate to themselves productivity improvements in detriment to the rest of society. In this sense, as argued by Booth (1995), the union wage dierential over nonunion alternatives is expected to be increasing with union power. In this paper we examine the role of unions in bringing about interindustry wage dierentials. Previous accounts on the role of unions in the process of wage determination in Brazil have focused on: (a) the growth of insider power and the decreasing importance of the government in collective bargaining (Carneiro and Henley, 1998); (b) the temporal stability of the interindustry wage structure (Gatica, Mizala and Romaguera, 1995); and (c) the role of human capital factors in explaining wage dierentials (Arbache, 1997). The assessment of how trade unions contribute to overall wage dispersion in Brazil, however, is yet to be done. Unlike previous ®ndings (Gosling and Machin, 1995; Fortin and Lemieux, 1997), our results indicate that under the intermediate centralized bargaining structure present in Brazil2 trade unions contribute to increasing rather than decreasing wage dispersion within the union sector. The paper is structured as follows. In the next section, we present the theoretical background underlying our analysis of union wage dierentials. In Section 3 we discuss the data used in this paper. In Section 4 we analyze several features of the data, exploiting the relevance of union power to explain the existence of a union wage dierential. Section 5 concludes the paper and discusses some of the possible implications of the results. 2. THEORETICAL BACKGROUND Before considering the empirical evidence, we will ®rst discuss how unions can have an impact upon wage determination and the consequences of dierent bargaining structures in shaping dierent outcomes in the labor market. A recent development in the economics of the labor market was the appearance during the 1980s of a wide range of studies on the behavior of trade unions and their impact on the process
of wage and employment determination. In the speci®c case of wage determination, it has been now widely suggested that wage moderation is usually associated with a more coordinated bargaining structure, or what has been termed corporatist arrangements. On the other extreme, however, explosive wage demands have been usually associated both with a lack of coordination and with synchronization in wage bargaining. The discussion on the role of institutions and wage bargaining structures in determining wage moderation cannot be dissociated from the debate on labor market ¯exibility. In the competitive model of the labor market, for example, the balance of supply and demand in the whole market dictates wages. Hicks's Theory of Wages (1932) was the starting point for this argument, but even Hicks himself became skeptical about the overall applicability of the competitive model, as he made clear in the preface to Hicks (1963). There has been much debate whether wages can be considered as ®xed by noncompetitive pressures. A common argument often used is that in most cases wages are determined by collective bargaining and also that other noncompetitive factors might be important (e.g., eciency wage and rent-sharing theories). The seminal contribution on the debate concerning the ability of organized groups to aect economic performance is attributed to Olson (1965, 1982). His view is that the ®nal eect of the action of interest groups in a society is mostly negative. Olson is particularly concerned with the fact that interest groups tend to act in detriment to the rest of the society by appropriating any improvement in eciency that their collective action might have generated. The accumulation of these groups may ultimately increase the complexity of regulation, the role of the government, reducing allocative eciency and thus the rate of economic growth. His preferred solution, therefore, seems to be a reduction in the capacity of economic interests to organize at all. Crouch (1985) argues that Olson's perception leads to a paradox. In Olson's view, once economic interests have become organized, they will be less likely to act in line with the idea of allocative eciency. But as the level of organization increases these special-interest groups may broaden their interests and become more politicized, aggregated and centralized. These encompassing organizations may then include concerns about economic growth among other
UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS
interests that they might also have (e.g., regarding the politics of income distribution). Considering the case of trade unions, Crouch suggests that the more centrally united is organized labor, the more are its actions compatible with the stability of the market economy (see also Bruno and Sachs, 1985). By investigating if, among the several characteristics that make up corporatism, centralization was empirically relevant, Calmfors and Drill (1988) were able to suggest that the claimed monotonic relationship between centralization and economic performance was not always true. Their results implied the existence of a hump-shaped relationship between unemployment and bargaining centralization. In the polar case of decentralization, negotiations between workers and ®rms are the ultimate determinants of wages; the fact that agents are too small in this context to interfere in the operation of the market means that wage restraint holds because competitive pressures put a cap on price and wage increases. At the other extreme of centralized, economy-wide bargaining, the idea is that corporatist arrangements may serve to increase the awareness of wage setters to the macroeconomic consequences of their actions. In periods of adverse shocks, corporatist institutions may thus contribute to reduce uncertainty regarding the deleterious consequences of possible distributional con¯ict and facilitate a relatively painless adjustment process. In the case of the intermediate group, where bargaining takes place mostly at the industry level, the explanation for the poorer performance relative to the two polar cases is usually the existence of poorly coordinated monopolistic power in the labor market that constrains the successful operation of either competitive market forces or corporatist coordination. In this context, trade unions' bargaining power may be quite strong since rapid labor turnover is restricted by high training costs and or hiring and ®ring costs. Insider bargaining power is also high, meaning that workers perceive that by pressing for higher wages the ®rm's product price may increase without however aecting the aggregate price level and therefore their real take home pay. As bargaining is at the industry level, an individual ®rm will not achieve a competitive advantage over the others and, assuming that industry demand is relatively inelastic, all the ®rms believe that the overall employment consequences will be insigni®cant. Overall, therefore, the situation of intermediate
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collective bargaining with strong trade unions may be quite conducive to poor employment and in¯ation performance, with rent sharing being a pervasive characteristic. Freeman (1988) and Rowthorn (1992) show that wage dispersion also tends to be lower in corporatist economies as opposed to economies with a decentralized labor market. In the case of an economy with an intermediate centralized bargaining structure, one should thus expect greater dispersion relatively to the polar cases. As the union wage dierential tends to increase with union power, the intermediate centralized bargaining structure seems to favor the existence of a rather large wage dispersion. In what follows, we present some evidence on the case of Brazil with the aim of assessing the validity of this argument. 3. DATA DESCRIPTION The micro-data used in our investigation are from the National Household Surveys (Pesquisa Nacional por Amostragem Domiciliar ± PNAD) of 1992 and 1995, which are conducted by the Brazilian Institute of Geography and Statistics (IBGE). The sample we use is ®ltered to include economically active individuals, nonemployers, aged between 18 and 65 years, who work in their main occupation, with formal labor contract, and are aliated to any of the 22 industries that compose the manufacturing sector. In Table 1 we present the means for a number of variables in both subsamples. Unionized workers seem to accumulate more human capital characteristics as illustrated by their means regarding education, tenure and experience. The subsample of nonunionized workers is made up of relatively more females and nonwhites, and who are paid less nonwage bene®ts. Nonunion workers are also likely to be found in the poorest regions, occupying low and medium-skill occupations, and are subject to more overtime work. Table 2 presents data on union density in the manufacturing sector. In general, union density varies from industry to industry. Transport material, mechanic, metallurgic, electronics, paperÐcapital-intensive industriesÐ present union densities above the weighted mean. On the other hand, traditionally laborintensive industries such as apparel, furniture, nonmetallic, and food present union densities below the weighted mean.
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WORLD DEVELOPMENT Table 1. Means for selected variables for union and nonunion workers
Variables
1992
1995
Unions
Nonunions
Unions
Nonunions
7.8 19.88 520.41 6.34 0.2281 0.7164 0.6666 0.6434 0.3343 0.2523
7.25 18.36 468.54 4.05 0.281 0.6025 0.564 0.6034 0.328 0.3123
8.41 19.5 499.27 6.61 0.2191 0.7069 0.6665 0.6672 0.3323 0.2437
7.66 18.42 471.16 4.11 0.2721 0.6077 0.5706 0.6046 0.3265 0.3154
Occupation Manager Professionals Clerical Sales Skilled a Semi-skilled a Unskilled a ( 1)
0.0787 0.0476 0.0898 0.0232 0.0708 0.5591 0.1310
0.0701 0.0336 0.1010 0.0467 0.0395 0.5363 0.1728
0.0828 0.0650 0.0833 0.0236 0.0701 0.5592 0.1161
0.0766 0.0317 0.0894 0.0543 0.0404 0.5349 0.1727
Region South Southeast North Centrewest Northeast ( 1)
0.3347 0.4626 0.0348 0.0219 0.1460
0.2551 0.4948 0.034 0.0502 0.1658
0.3079 0.4776 0.0295 0.0219 0.1631
0.2483 0.4957 0.0411 0.0596 0.1553
Non-wage bene®ts Housing Meals Transportation Education/training Health N
0.0330 0.5110 0.5544 0.0597 0.4128 4.0550
0.0525 0.3606 0.4765 0.0221 0.2523 6.7300
0.0260 0.6173 0.6032 0.0772 0.5078 4.1080
0.0388 0.4552 0.5187 0.0308 0.2843 6.8110
Education (years) Experience (years) Experience square Tenure (years) Gender (female 1) Married (married 1) Head of family ( 1) Race (white 1) Metropilitan area ( 1) Ovetime worked (48 h/week 1)
a Skilled craftsmen, foremen and kindred workers; Semi-Skilled operatives and kindred workers; and Unskilled service workers.
According to Lipietz (1997), there are some stylized facts indicating a relationship between ®rm size and union density. In large plants, where Fordist and Taylorist managerial procedures are usually applied, workers tend to be more organized and unionism tends to ¯ourish. In order to investigate this proposition, we have calculated the Pearson correlation coecient (two-tailed) between the union density and industry ®rm size mean.3 The estimated coecient for 1992 is 0.6244 (p 0.002), thus suggesting that union density is related to ®rm size. So far, therefore, the two subsamples provide evidence of some distinct characteristics for unionized and nonunionized workers. An interesting question at this point is whether
wage dierences between union and nonunion workers can be attributable not to unions, but to dierences in observable productive traits, occupation, and regional characteristics. In this case, unions would not aect wage determination and would not be responsible for the observable wage dierences. In order to assess this point, the next section investigates wage dierentials for both subsamples. 4. UNIONS AND WAGE DIFFERENTIALS The methodology we adopt to estimate wage dierentials follows the method proposed by Haisken-DeNew and Schmidt (1997), that improves the standard procedure popularized
UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS
correct standard error in a single regression step. The standard deviation of wage dierentials is computed as q uj : SD u n0 H uj uj ÿ n0 D V 3
Table 2. Union density Industry
1992
1995
Apparel Beverages Chemical Electronic Food Furniture Leather Mechanic Metallurgic Mineral Nonmetallic Other Paper Perfumes Pharmaceutical Plastic Publishing Rubber Textiles Tobacco Transport Wood Weighted mean
33.2 34.5 38.2 45.0 30.3 25.4 39.7 43.8 44.3 44.9 28.6 29.9 47.5 19.5 33.6 31.7 36.2 32.1 50.5 63.9 51.0 28.6 37.6
33.6 37.7 37.4 42.1 29.3 20.3 43.6 44.3 45.9 47.8 28.7 30.2 52.1 26.1 37.1 31.0 34.4 46.8 55.7 45.7 51.9 21.5 37.6
SD u gives the weighted and adjusted standard deviation of coecients, H (.) transforms a column vector into a diagonal matrix whose diagonal is given by the column vector itself, D denotes the column vector formed by the diagonal elements of a matrix, and V is the variance-covariance matrix. (a) Industry eects and union eects
by Krueger and Summers (1988).4 The wage equations are estimated in the following form: ln wij a bXi uZj eij ;
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1
where w is the natural logarithm of the hourly real wage of worker i in industry j, X is the vector of personal characteristics, occupations and regions, Z is the vector of industry dummies which includes all industries, a is the intercept term, e is a random disturbance term re¯ecting unobserved characteristics and the inherent randomness of earnings statistics, and b and u are the vectors of parameters to be estimated. Since in this model the crossproduct matrix of the regressors is not of full rank, a linear restriction is imposed on the us as follows, X uj nj 0; 2 where n is the employment share in industry j. The reported coecients are interpreted as the dierence between the mean log wage for a worker in industry j and the employment-share weighted mean log wage for workers in the entire sample (measured as the employmentshare weighted mean of the coecients on all industry dummy variables). The formulation given by Eqns. (1) and (2) provides both economically sensible coecients and their
Table 3 presents coecient estimates of interindustry wage dierentials and the calculated standard deviation of wage spread. After controls for human capital, demographic characteristics, occupations, regions and nonwage bene®ts have been applied, we can still observe some wage dispersion for both unionized and nonunionized workers.5 Capital-intensive industries6 tend to pay a wage premium, while traditional industries tend to pay lower wages for workers with apparently the same observable productive attainments and other characteristics as those in the capital-intensive industries. Despite the dierences in the characteristics of unionized and nonunion workers, the pro®le of wage dierentials seems to be similar across the groups. The estimated Pearson correlation coecients are 0.7796 (p 0.000) for 1992, and 0.7094 (p 0.000) for 1995, suggesting that, irrespective of the union status, there is an industry eect in wage determination; i.e., wage premia are paid to all workers regardless of their aliation to a trade union. In line with Pencavel (1991), this result might be viewed as the case in which union activism provokes spillovers over other sectors of the economy. Nonunion sectors would try to mimic the wage practices in the unionized sectors by fearing turnover costs, bad morale, and even the threat of unionization (Dickens, 1986). When we investigated this result further by running wage equations based on an unsplit sample we found a signi®cant union eect in manufacturing wage determination. Actually, the union dummy was statistically signi®cant for both years (0.1070; t-value 10.12, for 1992; and 0.0656, t-value 6.39 for 1995), suggesting that despite the common structure
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WORLD DEVELOPMENT Table 3. Wage dierentials for union and nonunion workers (standard errors in parentheses) 1992 Union
Apparel ÿ0.2081 Beverages ÿ0.0462 Chemical 0.3162 Electronic 0.1189 Food ÿ0.1385 Furniture ÿ0.2217 Leather ÿ0.3132 Mechanic 0.0462 Metallurgic 0.0553 Mineral 0.2445 Nonmetallic 0.0120 Other ÿ0.0861 Paper 0.0888 Perfumes 0.0769 Pharmaceutical 0.1612 Plastic ÿ0.0582 Publishing ÿ0.0214 Rubber 0.0458 Textiles ÿ0.1018 Tobacco 0.1215 Transport 0.2074 Wood ÿ0.1210 0.5582 R2 F 110.09 SD 0.1329 n 4.055
(0.0260) (0.0536) (0.0323) (0.0360) (0.0214) (0.0546) (0.0695) (0.0312) (0.0203) (0.0444) (0.0384) (0.0659) (0.0479) (0.1272) (0.0831) (0.0525) (0.0400) (0.0845) (0.0270) (0.0814) (0.0293) (0.0510)
a
1995 Nonunion
ÿ0.0959 ÿ0.0514 0.1228 0.0925 ÿ0.1080 ÿ0.0654 ÿ0.0710 0.0837 0.1146 0.0557 ÿ0.0073 ÿ0.0664 0.1005 0.0255 0.1109 0.0091 0.0266 0.0101 ÿ0.0122 0.2701 0.2561 ÿ0.1324 0.4959 142.94 0.0557 6.730
Union
(0.0178) (0.0381) (0.0247) (0.0328) (0.0134) (0.0311) (0.0553) (0.0272) (0.0180) (0.0391) (0.0238) (0.0420) (0.0444) (0.0610) (0.0576) (0.0349) (0.0294) (0.0568) (0.0270) (0.1063) (0.0291) (0.0317)
ÿ0.1502 ÿ0.0067 0.1589 0.0265 ÿ0.1495 ÿ0.0666 ÿ0.2179 0.0430 0.0498 0.2107 ÿ0.0886 ÿ0.0124 ÿ0.0132 0.1552 0.0379 ÿ0.1274 0.1520 0.0921 ÿ0.1250 0.1293 0.2374 ÿ0.0943 0.6605 209.36 0.105 4.108
(0.0249) (0.0480) (0.0327) (0.0345) (0.0204) (0.0592) (0.0757) (0.0288) (0.0192) (0.0409) (0.0357) (0.0530) (0.0452) (0.1011) (0.0819) (0.0517) (0.0400) (0.0628) (0.0279) (0.1057) (0.0255) (0.0542)
Nonunion ÿ0.0926 ÿ0.0026 0.0643 0.0605 ÿ0.0750 ÿ0.0710 ÿ0.1772 0.0640 0.0791 0.0488 ÿ0.0222 ÿ0.0265 0.0422 ÿ0.0270 0.2154 0.0153 0.0761 0.1020 0.0119 0.0817 0.1555 ÿ0.0706 0.6157 235.63 0.0099 6.811
(0.0175) (0.0372) (0.0249) (0.0295) (0.0127) (0.0294) (0.0663) (0.0256) (0.0178) (0.0385) (0.0225) (0.0345) (0.0471) (0.0595) (0.0623) (0.0344) (0.0284) (0.0587) (0.0314) (0.0965) (0.0262) (0.0281)
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, overtime work, and ®ve regional dummies (Northeast 1).
of union and nonunion wages, unions have a non-negligible role in wage determination. By their anti-logs, these coecients show that the union wage dierential in Brazil was of roughly 11.3% in 1992, and 6.7% in 1995.7 These ®gures are quite close to the international evidence surveyed by the World Bank (1995), p. 81). In developing countries like Mexico and Malaysia, the union wage union premium is estimated in the range of 10±20%. Similar ®gures are found for developed countries as well, as surveyed by Booth (1995), chapter 6). The union wage premium in Germany and the United Kingdom, for example, was estimated at 10% and 5%, respectively, during 1985±87. Theoretical justi®cations for this result can be searched in the monitoring and turnover models of eciency wages (see Shapiro and Stiglitz, 1984, and Stiglitz, 1985, respectively), which do not distinguish the union status of workers, or in normative eciency wage models that predict
the importance of employers' ability-to-pay (Akerlof, 1984). In order to assess the relevance of trade unions in aecting wage dierentials' we have estimated the weighted and adjusted standard deviation of wage dierentials.8 The results at the bottom of Table 3 show that: (i) the wage dispersion coecient found for unionized workers is greater than that found for nonunionized workers, and (ii) the relative dierence between the measures of dispersion increases over 1992±95. This stresses the point that trade unions cause greater dispersion of wages in the case of Brazil; a result which holds even under changes in the economic context with greater economic openness, lower in¯ation and increased competitiveness after 1994. Variations in the observed union wage gap across sectors are likely to be related to dierences in the surplus made available by ®rms, and to the ability of trade unions to appropri-
UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS
ate some of these rents. The varying bargaining power of dierent trade unions ampli®es wage dispersion for workers with apparently the same characteristics. (b) Further testing on union power In this section, we exploit further the data using the Blinder±Oaxaca decomposition methodology (see Blinder, 1973, and Oaxaca, 1973) to distinguish between the portion of wage dierences that is due to observable characteristics and the portion that is not. As in Booth (1995), we assume that the unexplained wage gap is due to union power. De®ne raw wage dierentials as U ÿ ln w N , where w is the average wage D ln w of U, unionized workers, and N, nonunionized workers. Considering the wage equations estimated so far, we compute the following: X U N DI aU ÿ aN lU X ÿ X X N 4 X lU ÿ lN ; X U N DII aU ÿ aN lN X ÿ X X U 5 X lU ÿ lN ; where a is the intercept, X the vector of exogenous variables for each subsample9 and l the vector of estimated coecients. In the absence of a union eectÐwhich is given by dierent payos to productive traitsÐwage dierentials would be explained only by dierences in mean values of human capital and other variables.10 Table 4 reports the results of such decomposition analysis. In 1992, roughly 29.6% (calculated as (0.1381 ÿ 0.0387)/0.3357), while in 1995 approximately 18.8% of raw wage dierentials between unionized and nonunionized workers are not explained by dierences in observable productive endowments, occupation and other variables,11 con®rming our previous ®ndings that unions have an important role in wage determination. More impor-
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tantly, however, these results state that, all else held constant, when a worker changes his/her union membership status from nonunion to union member s/he is liable to perceive a wage increase. It should be noted that our results contradict recent empirical evidence that attribute to trade unions the capacity to reduce rather than increasing wage dispersion within the union sector (e.g., Stewart, 1995; Gosling and Machin, 1995; Fortin and Lemieux, 1997). The ability to reduce wage dispersion seems to be more apparent in contexts in which labor markets operate freer of institutional rigidities (Freeman, 1988; Rowthorn, 1992). In cases where bargaining takes place at an intermediate centralized level, there is a considerable dierence in the bargaining power of dierent unions and this is believed to favor an increasing, rather than a decreasing, spread of earnings within the union sector. Our results, therefore, appear compatible with the structure of manufacturing labor markets in Brazil. 5. CONCLUSIONS In this paper, we have investigated the importance of trade unions in collective bargaining in the context of a developing country manufacturing labor market. Our ®ndings indicate that, in the case of Brazil, unions contribute to increasing rather than decreasing the spread of wages within the union sector. Moreover, for the case of Brazil, a change in membership status for manufacturing workers seems to be relevant, as unionized workers tend to earn higher wages. Previous studies on the union-nonunion wage dierential in developed countries, particularly the United Kingdom and the United States, have found lower dispersion within the union sector. In line with the earlier theoretical discussion, we have argued that the evidence presented here for the case of Brazil may be related to the structure of
Table 4. Decomposition analysis of wage dierences
a
Year
Dierential
Intercept
Method
Endowments
Coecients
1992
0.3357
0.1381
1995
0.3575
0.1895
I II I II
0.2363 0.233 0.2902 0.2935
ÿ0.0387 ÿ0.0354 ÿ0.1222 ÿ0.1255
a
Estimations based on means and coecients respectively from Tables 1 and 3.
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WORLD DEVELOPMENT
collective bargaining prevailing in Brazilian manufacturing. More speci®cally, we have found a positive correlation between ®rm size and union density in Brazil, and also that wage premia are paid to all workers irrespective of their aliation to a trade union. Further investigation of the data revealed however that, despite the apparently common structure of union and nonunion
wages, unions play a non-negligible role in bringing about interindustry wage dierentials. Our estimations show that wage dispersion is greater for unionized workers and that the relative dierence between the measures of dispersion increases over 1992±95. Furthermore, our results state that, in Brazil, when a worker becomes aliated to a trade union s/he is liable to perceive a wage increase.
NOTES 1. According to the literature that has been popularized by Calmfors and Drill (1988), an intermediate centralized level of collective bargaining occurs when bargaining takes place mostly at the industry level.
indeed a possibility. In the case of Brazilian manufacturing, however, Arbache (1999) shows that these results hold for a sample of male workers and also for a sample of female workers.
2. A discussion of Brazilian wage setting institutions can be found in Carneiro (1998) and Carneiro and Henley (1998).
6. The following industries are usually considered as capital-intensive: chemicals, electronic, mechanic, metallurgic, pharmaceutical, transport.
3. The data on industry ®rm size mean are from Relat orio Anual de Informacßo~es Sociais (RAIS), conducted by the Ministry of Labour. We have not computed correlation coecients for 1995 because the data were not available for that year.
7. The drop in wage dierentials 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.
4. Haisken-DeNew and Schmidt show that the Krueger and Summers procedure overstates substantially the standard error of coecients, 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 dierentials vector. In addition, the estimated variance of the renormalized coecients is overestimated in their procedure, which aects 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 dier signi®cantly by gender, and that is
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 eects not captured by our dataset may also aect this result, which can be thus either over or underestimated.
REFERENCES Akerlof, G. A. (1984) Gift exchange and eciency-wage theories: Four views. American Economic Review 74, 79±83. Allen, S. (1995) Update notes on the inter-industry wage structure: 1890±1990. Industrial and Labor Relations Review 48, 305±321. Arbache, J. S. (1997) Wage dierentials in Brazilian manufacturing. Studies in Economics 97/5, Depart-
ment of Economics, University of Kent at Canterbury, UK. Arbache, J. S. (1999) Do unions always decrease wage dispersion? The case of Brazilian manufacturing. Journal of Labor Research. Blanch¯ower, D. G., Oswald, A. J. and Sanfey, P. (1996) Wages, pro®ts and rent-sharing. Quarterly Journal of Economics 111, 227±253.
UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS Blinder, A. S. (1973) Wage discrimination: Reduced form and structural estimates. Journal of Human Resources 18, 436±455. Booth, A. L. (1995) The Economics of the Trade Union. Cambridge University Press, Cambridge. Bruno, M. and Sachs, J. (1985) The Economics of Worldwide Stag¯ation. Oxford, Basil Blackwell. Calmfors, L. and Drill, J. (1988) Bargaining structure, corporatism and macroeconomic performance. Economic Policy 6, 13±61. Carneiro, F. G. (1998) Productivity eects in Brazilian wage determination. World Development 26 (1), 139± 153. Carneiro, F.G. and Henley, A. (1998) Wage determination in Brazil: The growth of union bargaining power and informal employment. Journal of Development Studies 34 (4), 117±138. Crouch, C. (1985) Conditions for trade union wage restraint. In The Politics of In¯ation and Economic Stagnation: Theoretical Approaches and International Case Studies, eds. L. N. Lindberg, C. S. Maier, The Brookings Institution, Washington DC. Dickens, W. T. (1986) Wages, employment and the threat of collective action by workers. Working Paper No. 1856, National Bureau of Economic Research. Cambridge, MA. Fortin, N. M. and Lemieux, T. (1997) Institutional changes and rising wage inequality: Is there a linkage?. Journal of Economic Perspectives 11, 75±96. Freeman, R. (1988) Labor market institutions and economic performance. Economic Policy 6, 64±80. Gatica, J., Mizala, A. and Romaguera, P. (1995) Interindustry wage dierentials in Brazil. Economic Development and Cultural Change 43, 315±331. Gelb, A., Knight, J. B. and Sabot, R. H. (1991) Public sector employment, rent seeking and economic growth. Economic Journal 101, 1186±1199. Gosling, A. and Machin, S. (1995) Trade unions and the dispersion of earnings in British establishments, 1980±90. Oxford Bulletin of Economics and Statistics 57, 167±184. Haisken-DeNew, J. P. and Schmidt, C. M. (1997) Interindustry and inter-region dierentials: Mechanics
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and interpretation. Review of Economics and Statistics 79, 516±521. Hicks, R. (1963) The Theory of Wages. (1st edn. 1932), Macmillan, London. Holmlund, B. and Zetterberg, J. (1991) Insider eects in wage determination: Evidence from ®ve countries. European Economic Review 35 (1), 1009±1034. Krueger, A. B. and Summers, L. H. (1988) Eciency wages and the inter-industry wage structure. Econometrica 56, 259±193. Layard, R. and Nickell, S. J. (1986) Unemployment in the UK. Economica, Supplement 53, S121±S166. Lipietz, A. (1997) The post-fordist world: Labor relations, international hierarchy and global ecology. Review of International Political Economy 4, 1±41. Oaxaca, R. (1973) Male female wage dierentials in urban labor markets. International Economic Review 14, 693±709. Olson, M. (1965) The Logic of Collective Action. Harvard University Press, Cambridge, MA. Olson, M. (1982) The Rise and Decline of Nations. Yale University Press, New Haven, CT. Pencavel, J. (1991) Labor Markets under Trade Unionism. Blackwell, Oxford. Rowthorn, R. E. (1992) Centralization, employment and wage dispersion. Economic Journal 102, 506±523. Shapiro, C. and Stiglitz, J. E. (1984) Equilibrium unemployment as a worker discipline device. American Economic Review 74, 433±444. Slichter, S. (1950) Notes on the structure of wages. Review of Economics and Statistics 32, 80±91. Stewart, M. B. (1995) Union wage dierentials in an era of declining unionization. Oxford Bulletin of Economics and Statistics 57, 143±166. Stiglitz, J. (1985) Equilibrium wage distributions. Economic Journal 95, 595±618. Teal, F. (1996) The size and sources of economic rents in a developing country manufacturing labor market. Economic Journal 106, 963±976. World Bank (1995) Workers in an Integrating World: 1995 World Development Report. Oxford University Press, New York.
World Development Vol. 27, No. 10, pp. 1875±1883, 1999 Ó 1999 Elsevier Science Ltd. All rights reserved Printed in Great Britain 0305-750X/99/$ - see front matter
PII: S0305-750X(99)00090-X
Unions and Interindustry Wage Dierentials JORGE SABA ARBACHE Universidade 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 the
context of a developing country manufacturing labor market. The methodology we adopt to estimate wage dierentials follows the method proposed by Haisken-DeNew, J. P. and Schmidt, C. M. (1997) Review of Economics and Statistics 79, 516±521, since it improves 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 far greater in the unionized sector of Brazilian manufacturing, in contrast to evidence from other countries. Ó 1999 Elsevier Science Ltd. All rights reserved.
1. INTRODUCTION The structure of interindustry wages was the focus of intense research in the early 1950s and has received a great deal of attention in recent years. Some of the stylized facts found by earlier studies include evidence of large dierentials in wages across industries, even after controls have been implemented (Slichter, 1950), and a remarkable temporal persistence of the interindustry wage structure (e.g., Allen, 1995). Many authors have tried to argue that these stylized facts suce to rule out the adequacy of simple competitive models of the labor market against the superiority of alternative approaches of noncompetitive models such as eciency 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 focused attention on the evidence of a union-nonunion wage dierential (e.g., Booth, 1995). Models of the aggregate labor market, for example, have used the evidence of such a dierential as a proxy for union power. The underlying feature seems to be the existence of economic rents, which can be bargained for. Higher union wages re¯ect the ability of trade unions to force ®rms to give up some of their surplus. In this
sense, the union wage dierential would be positively related to union power, as assumed in aggregate labor market models (Layard and Nickell, 1986). In the study of developed country labor markets, the sources of these rents are usually associated with the size of the ®rms, industry aliation, and productive performance (e.g., Blanch¯ower, Oswald and Sanfey, 1996). For developing countries, however, there are dierent accounts on the source of these rents. Teal (1996), for example, suggests that the public sector would be the primary source of rents in developing country labor markets because of their characteristic of strong institutional in¯uence in wage determination. The public sector would be in general determining an overall wage policy as a ¯oor to wage negotiations and also establishing a distinction within the own public sector between productive and unproductive activities (Gelb, Knight and Sabot, 1991).
* 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.
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There is also some evidence that liberalization of collective bargaining in the context of a developing country facing an intermediate centralized bargaining structure1 may be rather conducive to rent sharing (Carneiro, 1998). Powerful wage bargainers would face little resistance to convert positive demand shocks into wage gains and also to appropriate to themselves productivity improvements in detriment to the rest of society. In this sense, as argued by Booth (1995), the union wage dierential over nonunion alternatives is expected to be increasing with union power. In this paper we examine the role of unions in bringing about interindustry wage dierentials. Previous accounts on the role of unions in the process of wage determination in Brazil have focused on: (a) the growth of insider power and the decreasing importance of the government in collective bargaining (Carneiro and Henley, 1998); (b) the temporal stability of the interindustry wage structure (Gatica, Mizala and Romaguera, 1995); and (c) the role of human capital factors in explaining wage dierentials (Arbache, 1997). The assessment of how trade unions contribute to overall wage dispersion in Brazil, however, is yet to be done. Unlike previous ®ndings (Gosling and Machin, 1995; Fortin and Lemieux, 1997), our results indicate that under the intermediate centralized bargaining structure present in Brazil2 trade unions contribute to increasing rather than decreasing wage dispersion within the union sector. The paper is structured as follows. In the next section, we present the theoretical background underlying our analysis of union wage dierentials. In Section 3 we discuss the data used in this paper. In Section 4 we analyze several features of the data, exploiting the relevance of union power to explain the existence of a union wage dierential. Section 5 concludes the paper and discusses some of the possible implications of the results. 2. THEORETICAL BACKGROUND Before considering the empirical evidence, we will ®rst discuss how unions can have an impact upon wage determination and the consequences of dierent bargaining structures in shaping dierent outcomes in the labor market. A recent development in the economics of the labor market was the appearance during the 1980s of a wide range of studies on the behavior of trade unions and their impact on the process
of wage and employment determination. In the speci®c case of wage determination, it has been now widely suggested that wage moderation is usually associated with a more coordinated bargaining structure, or what has been termed corporatist arrangements. On the other extreme, however, explosive wage demands have been usually associated both with a lack of coordination and with synchronization in wage bargaining. The discussion on the role of institutions and wage bargaining structures in determining wage moderation cannot be dissociated from the debate on labor market ¯exibility. In the competitive model of the labor market, for example, the balance of supply and demand in the whole market dictates wages. Hicks's Theory of Wages (1932) was the starting point for this argument, but even Hicks himself became skeptical about the overall applicability of the competitive model, as he made clear in the preface to Hicks (1963). There has been much debate whether wages can be considered as ®xed by noncompetitive pressures. A common argument often used is that in most cases wages are determined by collective bargaining and also that other noncompetitive factors might be important (e.g., eciency wage and rent-sharing theories). The seminal contribution on the debate concerning the ability of organized groups to aect economic performance is attributed to Olson (1965, 1982). His view is that the ®nal eect of the action of interest groups in a society is mostly negative. Olson is particularly concerned with the fact that interest groups tend to act in detriment to the rest of the society by appropriating any improvement in eciency that their collective action might have generated. The accumulation of these groups may ultimately increase the complexity of regulation, the role of the government, reducing allocative eciency and thus the rate of economic growth. His preferred solution, therefore, seems to be a reduction in the capacity of economic interests to organize at all. Crouch (1985) argues that Olson's perception leads to a paradox. In Olson's view, once economic interests have become organized, they will be less likely to act in line with the idea of allocative eciency. But as the level of organization increases these special-interest groups may broaden their interests and become more politicized, aggregated and centralized. These encompassing organizations may then include concerns about economic growth among other
UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS
interests that they might also have (e.g., regarding the politics of income distribution). Considering the case of trade unions, Crouch suggests that the more centrally united is organized labor, the more are its actions compatible with the stability of the market economy (see also Bruno and Sachs, 1985). By investigating if, among the several characteristics that make up corporatism, centralization was empirically relevant, Calmfors and Drill (1988) were able to suggest that the claimed monotonic relationship between centralization and economic performance was not always true. Their results implied the existence of a hump-shaped relationship between unemployment and bargaining centralization. In the polar case of decentralization, negotiations between workers and ®rms are the ultimate determinants of wages; the fact that agents are too small in this context to interfere in the operation of the market means that wage restraint holds because competitive pressures put a cap on price and wage increases. At the other extreme of centralized, economy-wide bargaining, the idea is that corporatist arrangements may serve to increase the awareness of wage setters to the macroeconomic consequences of their actions. In periods of adverse shocks, corporatist institutions may thus contribute to reduce uncertainty regarding the deleterious consequences of possible distributional con¯ict and facilitate a relatively painless adjustment process. In the case of the intermediate group, where bargaining takes place mostly at the industry level, the explanation for the poorer performance relative to the two polar cases is usually the existence of poorly coordinated monopolistic power in the labor market that constrains the successful operation of either competitive market forces or corporatist coordination. In this context, trade unions' bargaining power may be quite strong since rapid labor turnover is restricted by high training costs and or hiring and ®ring costs. Insider bargaining power is also high, meaning that workers perceive that by pressing for higher wages the ®rm's product price may increase without however aecting the aggregate price level and therefore their real take home pay. As bargaining is at the industry level, an individual ®rm will not achieve a competitive advantage over the others and, assuming that industry demand is relatively inelastic, all the ®rms believe that the overall employment consequences will be insigni®cant. Overall, therefore, the situation of intermediate
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collective bargaining with strong trade unions may be quite conducive to poor employment and in¯ation performance, with rent sharing being a pervasive characteristic. Freeman (1988) and Rowthorn (1992) show that wage dispersion also tends to be lower in corporatist economies as opposed to economies with a decentralized labor market. In the case of an economy with an intermediate centralized bargaining structure, one should thus expect greater dispersion relatively to the polar cases. As the union wage dierential tends to increase with union power, the intermediate centralized bargaining structure seems to favor the existence of a rather large wage dispersion. In what follows, we present some evidence on the case of Brazil with the aim of assessing the validity of this argument. 3. DATA DESCRIPTION The micro-data used in our investigation are from the National Household Surveys (Pesquisa Nacional por Amostragem Domiciliar ± PNAD) of 1992 and 1995, which are conducted by the Brazilian Institute of Geography and Statistics (IBGE). The sample we use is ®ltered to include economically active individuals, nonemployers, aged between 18 and 65 years, who work in their main occupation, with formal labor contract, and are aliated to any of the 22 industries that compose the manufacturing sector. In Table 1 we present the means for a number of variables in both subsamples. Unionized workers seem to accumulate more human capital characteristics as illustrated by their means regarding education, tenure and experience. The subsample of nonunionized workers is made up of relatively more females and nonwhites, and who are paid less nonwage bene®ts. Nonunion workers are also likely to be found in the poorest regions, occupying low and medium-skill occupations, and are subject to more overtime work. Table 2 presents data on union density in the manufacturing sector. In general, union density varies from industry to industry. Transport material, mechanic, metallurgic, electronics, paperÐcapital-intensive industriesÐ present union densities above the weighted mean. On the other hand, traditionally laborintensive industries such as apparel, furniture, nonmetallic, and food present union densities below the weighted mean.
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WORLD DEVELOPMENT Table 1. Means for selected variables for union and nonunion workers
Variables
1992
1995
Unions
Nonunions
Unions
Nonunions
7.8 19.88 520.41 6.34 0.2281 0.7164 0.6666 0.6434 0.3343 0.2523
7.25 18.36 468.54 4.05 0.281 0.6025 0.564 0.6034 0.328 0.3123
8.41 19.5 499.27 6.61 0.2191 0.7069 0.6665 0.6672 0.3323 0.2437
7.66 18.42 471.16 4.11 0.2721 0.6077 0.5706 0.6046 0.3265 0.3154
Occupation Manager Professionals Clerical Sales Skilled a Semi-skilled a Unskilled a ( 1)
0.0787 0.0476 0.0898 0.0232 0.0708 0.5591 0.1310
0.0701 0.0336 0.1010 0.0467 0.0395 0.5363 0.1728
0.0828 0.0650 0.0833 0.0236 0.0701 0.5592 0.1161
0.0766 0.0317 0.0894 0.0543 0.0404 0.5349 0.1727
Region South Southeast North Centrewest Northeast ( 1)
0.3347 0.4626 0.0348 0.0219 0.1460
0.2551 0.4948 0.034 0.0502 0.1658
0.3079 0.4776 0.0295 0.0219 0.1631
0.2483 0.4957 0.0411 0.0596 0.1553
Non-wage bene®ts Housing Meals Transportation Education/training Health N
0.0330 0.5110 0.5544 0.0597 0.4128 4.0550
0.0525 0.3606 0.4765 0.0221 0.2523 6.7300
0.0260 0.6173 0.6032 0.0772 0.5078 4.1080
0.0388 0.4552 0.5187 0.0308 0.2843 6.8110
Education (years) Experience (years) Experience square Tenure (years) Gender (female 1) Married (married 1) Head of family ( 1) Race (white 1) Metropilitan area ( 1) Ovetime worked (48 h/week 1)
a Skilled craftsmen, foremen and kindred workers; Semi-Skilled operatives and kindred workers; and Unskilled service workers.
According to Lipietz (1997), there are some stylized facts indicating a relationship between ®rm size and union density. In large plants, where Fordist and Taylorist managerial procedures are usually applied, workers tend to be more organized and unionism tends to ¯ourish. In order to investigate this proposition, we have calculated the Pearson correlation coecient (two-tailed) between the union density and industry ®rm size mean.3 The estimated coecient for 1992 is 0.6244 (p 0.002), thus suggesting that union density is related to ®rm size. So far, therefore, the two subsamples provide evidence of some distinct characteristics for unionized and nonunionized workers. An interesting question at this point is whether
wage dierences between union and nonunion workers can be attributable not to unions, but to dierences in observable productive traits, occupation, and regional characteristics. In this case, unions would not aect wage determination and would not be responsible for the observable wage dierences. In order to assess this point, the next section investigates wage dierentials for both subsamples. 4. UNIONS AND WAGE DIFFERENTIALS The methodology we adopt to estimate wage dierentials follows the method proposed by Haisken-DeNew and Schmidt (1997), that improves the standard procedure popularized
UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS
correct standard error in a single regression step. The standard deviation of wage dierentials is computed as q uj : SD u n0 H uj uj ÿ n0 D V 3
Table 2. Union density Industry
1992
1995
Apparel Beverages Chemical Electronic Food Furniture Leather Mechanic Metallurgic Mineral Nonmetallic Other Paper Perfumes Pharmaceutical Plastic Publishing Rubber Textiles Tobacco Transport Wood Weighted mean
33.2 34.5 38.2 45.0 30.3 25.4 39.7 43.8 44.3 44.9 28.6 29.9 47.5 19.5 33.6 31.7 36.2 32.1 50.5 63.9 51.0 28.6 37.6
33.6 37.7 37.4 42.1 29.3 20.3 43.6 44.3 45.9 47.8 28.7 30.2 52.1 26.1 37.1 31.0 34.4 46.8 55.7 45.7 51.9 21.5 37.6
SD u gives the weighted and adjusted standard deviation of coecients, H (.) transforms a column vector into a diagonal matrix whose diagonal is given by the column vector itself, D denotes the column vector formed by the diagonal elements of a matrix, and V is the variance-covariance matrix. (a) Industry eects and union eects
by Krueger and Summers (1988).4 The wage equations are estimated in the following form: ln wij a bXi uZj eij ;
1879
1
where w is the natural logarithm of the hourly real wage of worker i in industry j, X is the vector of personal characteristics, occupations and regions, Z is the vector of industry dummies which includes all industries, a is the intercept term, e is a random disturbance term re¯ecting unobserved characteristics and the inherent randomness of earnings statistics, and b and u are the vectors of parameters to be estimated. Since in this model the crossproduct matrix of the regressors is not of full rank, a linear restriction is imposed on the us as follows, X uj nj 0; 2 where n is the employment share in industry j. The reported coecients are interpreted as the dierence between the mean log wage for a worker in industry j and the employment-share weighted mean log wage for workers in the entire sample (measured as the employmentshare weighted mean of the coecients on all industry dummy variables). The formulation given by Eqns. (1) and (2) provides both economically sensible coecients and their
Table 3 presents coecient estimates of interindustry wage dierentials and the calculated standard deviation of wage spread. After controls for human capital, demographic characteristics, occupations, regions and nonwage bene®ts have been applied, we can still observe some wage dispersion for both unionized and nonunionized workers.5 Capital-intensive industries6 tend to pay a wage premium, while traditional industries tend to pay lower wages for workers with apparently the same observable productive attainments and other characteristics as those in the capital-intensive industries. Despite the dierences in the characteristics of unionized and nonunion workers, the pro®le of wage dierentials seems to be similar across the groups. The estimated Pearson correlation coecients are 0.7796 (p 0.000) for 1992, and 0.7094 (p 0.000) for 1995, suggesting that, irrespective of the union status, there is an industry eect in wage determination; i.e., wage premia are paid to all workers regardless of their aliation to a trade union. In line with Pencavel (1991), this result might be viewed as the case in which union activism provokes spillovers over other sectors of the economy. Nonunion sectors would try to mimic the wage practices in the unionized sectors by fearing turnover costs, bad morale, and even the threat of unionization (Dickens, 1986). When we investigated this result further by running wage equations based on an unsplit sample we found a signi®cant union eect in manufacturing wage determination. Actually, the union dummy was statistically signi®cant for both years (0.1070; t-value 10.12, for 1992; and 0.0656, t-value 6.39 for 1995), suggesting that despite the common structure
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WORLD DEVELOPMENT Table 3. Wage dierentials for union and nonunion workers (standard errors in parentheses) 1992 Union
Apparel ÿ0.2081 Beverages ÿ0.0462 Chemical 0.3162 Electronic 0.1189 Food ÿ0.1385 Furniture ÿ0.2217 Leather ÿ0.3132 Mechanic 0.0462 Metallurgic 0.0553 Mineral 0.2445 Nonmetallic 0.0120 Other ÿ0.0861 Paper 0.0888 Perfumes 0.0769 Pharmaceutical 0.1612 Plastic ÿ0.0582 Publishing ÿ0.0214 Rubber 0.0458 Textiles ÿ0.1018 Tobacco 0.1215 Transport 0.2074 Wood ÿ0.1210 0.5582 R2 F 110.09 SD 0.1329 n 4.055
(0.0260) (0.0536) (0.0323) (0.0360) (0.0214) (0.0546) (0.0695) (0.0312) (0.0203) (0.0444) (0.0384) (0.0659) (0.0479) (0.1272) (0.0831) (0.0525) (0.0400) (0.0845) (0.0270) (0.0814) (0.0293) (0.0510)
a
1995 Nonunion
ÿ0.0959 ÿ0.0514 0.1228 0.0925 ÿ0.1080 ÿ0.0654 ÿ0.0710 0.0837 0.1146 0.0557 ÿ0.0073 ÿ0.0664 0.1005 0.0255 0.1109 0.0091 0.0266 0.0101 ÿ0.0122 0.2701 0.2561 ÿ0.1324 0.4959 142.94 0.0557 6.730
Union
(0.0178) (0.0381) (0.0247) (0.0328) (0.0134) (0.0311) (0.0553) (0.0272) (0.0180) (0.0391) (0.0238) (0.0420) (0.0444) (0.0610) (0.0576) (0.0349) (0.0294) (0.0568) (0.0270) (0.1063) (0.0291) (0.0317)
ÿ0.1502 ÿ0.0067 0.1589 0.0265 ÿ0.1495 ÿ0.0666 ÿ0.2179 0.0430 0.0498 0.2107 ÿ0.0886 ÿ0.0124 ÿ0.0132 0.1552 0.0379 ÿ0.1274 0.1520 0.0921 ÿ0.1250 0.1293 0.2374 ÿ0.0943 0.6605 209.36 0.105 4.108
(0.0249) (0.0480) (0.0327) (0.0345) (0.0204) (0.0592) (0.0757) (0.0288) (0.0192) (0.0409) (0.0357) (0.0530) (0.0452) (0.1011) (0.0819) (0.0517) (0.0400) (0.0628) (0.0279) (0.1057) (0.0255) (0.0542)
Nonunion ÿ0.0926 ÿ0.0026 0.0643 0.0605 ÿ0.0750 ÿ0.0710 ÿ0.1772 0.0640 0.0791 0.0488 ÿ0.0222 ÿ0.0265 0.0422 ÿ0.0270 0.2154 0.0153 0.0761 0.1020 0.0119 0.0817 0.1555 ÿ0.0706 0.6157 235.63 0.0099 6.811
(0.0175) (0.0372) (0.0249) (0.0295) (0.0127) (0.0294) (0.0663) (0.0256) (0.0178) (0.0385) (0.0225) (0.0345) (0.0471) (0.0595) (0.0623) (0.0344) (0.0284) (0.0587) (0.0314) (0.0965) (0.0262) (0.0281)
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, overtime work, and ®ve regional dummies (Northeast 1).
of union and nonunion wages, unions have a non-negligible role in wage determination. By their anti-logs, these coecients show that the union wage dierential in Brazil was of roughly 11.3% in 1992, and 6.7% in 1995.7 These ®gures are quite close to the international evidence surveyed by the World Bank (1995), p. 81). In developing countries like Mexico and Malaysia, the union wage union premium is estimated in the range of 10±20%. Similar ®gures are found for developed countries as well, as surveyed by Booth (1995), chapter 6). The union wage premium in Germany and the United Kingdom, for example, was estimated at 10% and 5%, respectively, during 1985±87. Theoretical justi®cations for this result can be searched in the monitoring and turnover models of eciency wages (see Shapiro and Stiglitz, 1984, and Stiglitz, 1985, respectively), which do not distinguish the union status of workers, or in normative eciency wage models that predict
the importance of employers' ability-to-pay (Akerlof, 1984). In order to assess the relevance of trade unions in aecting wage dierentials' we have estimated the weighted and adjusted standard deviation of wage dierentials.8 The results at the bottom of Table 3 show that: (i) the wage dispersion coecient found for unionized workers is greater than that found for nonunionized workers, and (ii) the relative dierence between the measures of dispersion increases over 1992±95. This stresses the point that trade unions cause greater dispersion of wages in the case of Brazil; a result which holds even under changes in the economic context with greater economic openness, lower in¯ation and increased competitiveness after 1994. Variations in the observed union wage gap across sectors are likely to be related to dierences in the surplus made available by ®rms, and to the ability of trade unions to appropri-
UNIONS AND INTERINDUSTRY WAGE DIFFERENTIALS
ate some of these rents. The varying bargaining power of dierent trade unions ampli®es wage dispersion for workers with apparently the same characteristics. (b) Further testing on union power In this section, we exploit further the data using the Blinder±Oaxaca decomposition methodology (see Blinder, 1973, and Oaxaca, 1973) to distinguish between the portion of wage dierences that is due to observable characteristics and the portion that is not. As in Booth (1995), we assume that the unexplained wage gap is due to union power. De®ne raw wage dierentials as U ÿ ln w N , where w is the average wage D ln w of U, unionized workers, and N, nonunionized workers. Considering the wage equations estimated so far, we compute the following: X U N DI aU ÿ aN lU X ÿ X X N 4 X lU ÿ lN ; X U N DII aU ÿ aN lN X ÿ X X U 5 X lU ÿ lN ; where a is the intercept, X the vector of exogenous variables for each subsample9 and l the vector of estimated coecients. In the absence of a union eectÐwhich is given by dierent payos to productive traitsÐwage dierentials would be explained only by dierences in mean values of human capital and other variables.10 Table 4 reports the results of such decomposition analysis. In 1992, roughly 29.6% (calculated as (0.1381 ÿ 0.0387)/0.3357), while in 1995 approximately 18.8% of raw wage dierentials between unionized and nonunionized workers are not explained by dierences in observable productive endowments, occupation and other variables,11 con®rming our previous ®ndings that unions have an important role in wage determination. More impor-
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tantly, however, these results state that, all else held constant, when a worker changes his/her union membership status from nonunion to union member s/he is liable to perceive a wage increase. It should be noted that our results contradict recent empirical evidence that attribute to trade unions the capacity to reduce rather than increasing wage dispersion within the union sector (e.g., Stewart, 1995; Gosling and Machin, 1995; Fortin and Lemieux, 1997). The ability to reduce wage dispersion seems to be more apparent in contexts in which labor markets operate freer of institutional rigidities (Freeman, 1988; Rowthorn, 1992). In cases where bargaining takes place at an intermediate centralized level, there is a considerable dierence in the bargaining power of dierent unions and this is believed to favor an increasing, rather than a decreasing, spread of earnings within the union sector. Our results, therefore, appear compatible with the structure of manufacturing labor markets in Brazil. 5. CONCLUSIONS In this paper, we have investigated the importance of trade unions in collective bargaining in the context of a developing country manufacturing labor market. Our ®ndings indicate that, in the case of Brazil, unions contribute to increasing rather than decreasing the spread of wages within the union sector. Moreover, for the case of Brazil, a change in membership status for manufacturing workers seems to be relevant, as unionized workers tend to earn higher wages. Previous studies on the union-nonunion wage dierential in developed countries, particularly the United Kingdom and the United States, have found lower dispersion within the union sector. In line with the earlier theoretical discussion, we have argued that the evidence presented here for the case of Brazil may be related to the structure of
Table 4. Decomposition analysis of wage dierences
a
Year
Dierential
Intercept
Method
Endowments
Coecients
1992
0.3357
0.1381
1995
0.3575
0.1895
I II I II
0.2363 0.233 0.2902 0.2935
ÿ0.0387 ÿ0.0354 ÿ0.1222 ÿ0.1255
a
Estimations based on means and coecients respectively from Tables 1 and 3.
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WORLD DEVELOPMENT
collective bargaining prevailing in Brazilian manufacturing. More speci®cally, we have found a positive correlation between ®rm size and union density in Brazil, and also that wage premia are paid to all workers irrespective of their aliation to a trade union. Further investigation of the data revealed however that, despite the apparently common structure of union and nonunion
wages, unions play a non-negligible role in bringing about interindustry wage dierentials. Our estimations show that wage dispersion is greater for unionized workers and that the relative dierence between the measures of dispersion increases over 1992±95. Furthermore, our results state that, in Brazil, when a worker becomes aliated to a trade union s/he is liable to perceive a wage increase.
NOTES 1. According to the literature that has been popularized by Calmfors and Drill (1988), an intermediate centralized level of collective bargaining occurs when bargaining takes place mostly at the industry level.
indeed a possibility. In the case of Brazilian manufacturing, however, Arbache (1999) shows that these results hold for a sample of male workers and also for a sample of female workers.
2. A discussion of Brazilian wage setting institutions can be found in Carneiro (1998) and Carneiro and Henley (1998).
6. The following industries are usually considered as capital-intensive: chemicals, electronic, mechanic, metallurgic, pharmaceutical, transport.
3. The data on industry ®rm size mean are from Relat orio Anual de Informacßo~es Sociais (RAIS), conducted by the Ministry of Labour. We have not computed correlation coecients for 1995 because the data were not available for that year.
7. The drop in wage dierentials 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.
4. Haisken-DeNew and Schmidt show that the Krueger and Summers procedure overstates substantially the standard error of coecients, 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 dierentials vector. In addition, the estimated variance of the renormalized coecients is overestimated in their procedure, which aects 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 dier signi®cantly by gender, and that is
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 eects not captured by our dataset may also aect this result, which can be thus either over or underestimated.
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