What explains wage differences between union members and covered nonmembers?

• Author: Schumacher, Edward J.

1. Introduction

   Economists have devoted considerable attention to understanding the impact of unions on labor market outcomes, in particular to the estimation and interpretation of union/nonunion wage differentials.l Less attention has been devoted, however, to the relevant measure of unionization. A worker can be covered by a collective bargaining agreement yet not be a member of the union. Lewis (1986) reviews previous studies examining the use of union membership or contract coverage as the appropriate measure of unionization and concludes that, while the use of membership yields a slightly higher coefficient estimate, the difference is small.

   The primary piece of legislation concerning union security is the Taft-Hartley Act of 1947, which made closed shops illegal.(2) In addition, Section 14 (b) of the act allowed states to make it illegal to require union membership or the payment of dues as a condition of continued employment. Thus, many states, primarily in the southeast and mountain regions, enacted right-to-work (RTW) laws. Workers in RTW states have the option of joining their union or not joining and remaining covered by the union contract. The union must equally represent all covered employees regardless of their membership status with respect to contract terms and representation in the grievance process. The National Labor Relations Act does not require employees covered by union security clauses to join a union or to support union activities beyond those relevant to collective bargaining, contract administration, and grievance adjustment (NLRB v. General Motors Corp. 373 U.S. 734, 1963). The covered nonmember, however, does not have the right to vote on union contracts or participate in other decision-making aspects of the union. In non-right-to-work states, a new employee is required to join the union if there is a union security clause in place, typically before the end of an initial 90-day probationary period.

   There has been considerable research on the determinants and effects of right-to-work laws (see, e.g., Farber 1984; Ellwood and Fine 1987; or see Moore and Newman [1985] and Moore [1998] for reviews of the literature). To the extent that covered nonmembers enjoy the same benefits of representation as members but with lower union dues, individuals have an incentive to freeride on the union contract. The more freeriders there are, however, the less effective the union will be and the benefits of unionization will be eroded. One strand of research in this area has focused on this freerider aspect of covered nonmembers (Booth 1985; Naylor 1989; Chaison and Dhavale 1992; Davis and Huston 1993; Naylor and Cripps 1993; Sobel 1995). Alternatively, others studies have attempted to estimate and interpret the wage differential between covered members and nonmembers referred to here as a membership differential or freerider penalty. Early studies using cross-sectional data typically found small significant differentials. For example, Jones (1982) uses data from the 1971 panels of the National Longitudinal Survey (NLS) and finds a membership differential of about 2% for young men and 3% for young women, with larger differentials for black men than white men and for white females than black females. Christensen and Maki (1983) find similar membership wage differentials.

   More recent cross-sectional studies have found a larger membership wage differential. For example, Budd and Na (1997) find a membership premium of 11-14% over covered nonmembers when using ordinary least squares (OLS) and larger estimates when using instrumental variable or sample selection methods. This differential is larger for men than women, but there are few differences across race and marital status, while the member differential varies substantially across occupational groups. Blakemore, Hunt, and Kiker (1986) and Hunt, Kiker, and Williams (1987) also find relatively large union membership differentials (about 13%) over covered nonmembers.

   A recent paper by Reilly (1996), however, using Canadian establishment-level data, finds little evidence for a membership differential. Once the union has organized more than 25% of the work force, there are significant establishment-wide effects on wages but no difference within establishments. Only for low union density establishments is the union able to capture a premium for members. These results suggest there is no establishment level freerider penalty.

   This paper attempts to reconcile the seemingly conflicting findings between Reilly's conclusion that there is little evidence for a freerider penalty at the establishment level and those using nonestablishment cross-sectional data sets, which typically find significant freerider penalties. A common explanation consistent with both findings is provided. Evidence is first provided on the extent of the freerider penalty utilizing a large pooled time-series cross-sectional data set. Next, the paper examines the source of the differential. Freerider status appears to be correlated with the strength of the union. Consistent with the findings of Reilly (1996), when there are few freeriders in a particular occupation/industry sector, the union/nonunion differential is high, and when there is a high proportion of freeriders in a sector, there is only a small difference between union and nonunion wages. In addition, the impact of such factors as unmeasured ability, the probationary period of joining the union, and measurement error in the data are examined. None of these factors explains a sizable portion of the freerider penalty.

2. Explaining the Membership Wage Differential

   Despite the legal requirement that union contracts equally represent members and nonmembers, cross-sectional data may reveal a wage difference between these workers. One possible explanation is that freeriders proxy the strength of the union in a particular sector. Given the findings of Reilly (1996) that there is no freerider penalty at the establishment level, wage differentials across establishments must be capturing factors other than freeriding per se. Workers are more likely to join the union when the gains are high and have little incentive to join when benefits are low. Union membership in this case reflects stronger union wage effects in some bargaining units over others rather than a causal effect of membership on wages.

   In addition to the union power hypothesis, there are other possible explanations for the freerider differential. There may be differences in ability, tastes, or motivation between members and freeriders not accounted for in micro level data sets. If workers with a stronger attachment to the job or who have higher (unmeasured) ability are more likely to join the union, then the freerider penalty will capture the lower ability of these individuals. Since the membership variable is capturing higher ability workers, statistical methods that better control for ability, motivation, and so forth should cause the freerider penalty to erode.

   Second, the freerider penalty could be a result of the probationary period of initial union membership. In jobs with union shop contract clauses in non-RTW states, a newly hired worker is initially covered by the collective bargaining agreement but typically has a 90-day probationary period. In RTW states, a newly hired employee need not join, but in either case, freeriding is more likely early in the stages of a new job and may be a proxy for job tenure, a wage determinant not included in standard cross-sectional studies (Reilly's analysis includes both experience and tenure). Thus, the union membership differential may be driven by a higher level of tenure on the job for union members over covered nonmembers.

   The freerider penalty may also be due to a higher degree of misclassification in the data among covered nonmembers than members. The Current Population Survey (CPS) data used here are obtained from surveys, which first ask individuals if they are members of a labor union and, if not, asks if they are covered by a collective bargaining agreement. If individuals frequently report that they are covered nonmembers when in fact they are not covered, the data will reveal a wage differential for membership since this variable is a cleaner estimate of union representation.

3. Estimating the Union Freerider Penalty

   Measurement

   In order to examine the effects of union membership and contract coverage on wages, we start with a simple log wage equation of the form

   ln [W.sub.i] = [X.sub.ji][[Beta].sub.j] + [Gamma][UNION.sub.i] + [e.sub.i], (1)

   where In [W.sub.i] is the log real wage of worker i, vector X consists of variables (indexed by j) measuring personal and job-related characteristics, and [Beta] is the coefficient vector. UNION is a dummy variable indicating union status, which could be defined as either membership or coverage. The coefficient [Gamma] measures the union wage differential under the assumption that union and nonunion wages differ only in the intercept term. For now, e is assumed to be a well-behaved error term. Alternatively, the model could be specified as

   ln [W.sub.i] [X.sub.ji][[Beta].sub.j] + [Delta][COVERED .sub.i] + [Theta][FREERIDER.sub.i] + [e.sub.i], (2)

   where COVERED is a dummy variable equal to one if the individual is covered by a collective bargaining agreement and FREERIDER is equal to one if the individual is a covered nonmember. The coefficient estimate of [Delta] provides a measure of the impact of contract coverage on wages, and [Theta] provides a measure of the wage penalty associated with freeriding. Restricting the sample to those covered by a collective bargaining contract and including a single freerider dummy provides an estimate of the freerider differential without restricting the coefficient estimates on all the other right-hand side variables to be equal in and out of the union sector.

   Due to potential unmeasured (to the researcher) ability correlated with union...