Do unions affect faculty salaries?

Economics of Education Review, Vol. 11, No. 3. pp. 219-223. 1992. Printed in Great Britain.

0272-7757/92$5.00 + 0.00 @ 1992Pergamon Press Ltd

Do Unions Affect Faculty Salaries? JAVED

Department

ASHRAF

of Finance and Economics, University of West Florida, Pensacola, FL 32514, U.S.A.

Abstract-This article calculates the union relative wage effect for a large sample of college faculty in the United States. The union wage premium is found to be 4.40% on average, but varies from -7.63 to 13.07% for different faculty subgroups. Faculty characteristics are found to be more important in influencing salaries in nonunion schools than in unionized institutions. INTRODUCTION WHILE

NUMEROUS STUDIES have attempted to gauge the effects of unions on salaries, few have attempted to examine the impact of collective bargaining on faculty salaries. This is probably attributable to the recency of the collective bargaining movement in higher education. Kelley (1977) reports that there were only 13 faculty-unionized educational institutions in the United States in 1968. Within 10 years, however, 5.50 campuses were represented by collective bargaining units. The relatively short time for which many institutions have been organized has been an impediment in making assessments of the impact of unions on faculty salaries. In this article, an attempt has been made to examine how the union-nonunion salary differential varies across a large sample of college faculty in the United States. Union salary premiums for as many as 21 faculty subgroups within the sample have been calculated. Unlike most earlier studies which used highly aggregated data, this article used a large micro data set. Thus, professional characteristics of individual faculty members can be controlled for in assessing union salary effects.

THE DATA AND VARIABLES The data used in this Survey of the American 4250 faculty members at education in the United

USED

article are from The 1977 Professoriate, and cover 158 institutions of higher States. Although the data

are somewhat dated, no other publicly available micro database exists which would allow an examination of union effects on faculty salaries on a national basis. Usable information on union membership was not available from the data set. However, it was possible to determine which schools were organized in 1977 from the 1984 Directory of Faculty Contracts and Bargaining Agents in Institutions of Higher Education. ’ The dependent variable is the log of monthly salaries. (Annual salaries were divided by 9.5 if the respondent reported that his or her salary was earned over a 9/10 month period, and by 11.5 if the period over which it was earned was reported as 1 l/ 12 months.) EXPERIENCE was defined as the period since the respondent assumed his or her first full-time teaching position while YRSINRANK represented the number of years that a faculty member had served within a certain rank. Dummy variables were defined for Assistant, Associate and Full Professors (with Instructors being the omitted base variable). TENURED was a dummy variable that distinguished between tenured and non-tenured faculty; HVYLOAD was a dummy variable with a value of 1 for teaching loads of 9 or more hours a semester, and 0 for less than 9 hours; BOOKS and ARTICLES represented the number of books and articles published by each respondent; and RESEARCH was a dummy variable distinguishing between a primarily research-oriented and a primarily teaching-oriented faculty member.* PRIVATE was 1 for faculty associated with private

[Manuscript received 25 January 1991; revision accepted for publication 6 December 1991.1 219

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institutions and 0 for those at public schools; and DIVl, DIV2 and DIV3 represented different kinds of institutions as categorized by the AAUP.3 AGE, MALE and BLACK are self-explanatory. UNION refers to all faculty members associated with an institution that was represented by a collective bargaining agent. Union-nonunion salary differentials were calculated as A = exp (U) - 1, where U is the coefficient estimate of the UNION variable in the wage equation. Some researchers have recently used simultaneous equation models to compute union wage premiums. Such models have been criticized, among others, by Freeman and Medoff (1981) and Barbezat (1989). This study lists estimates from such a model in Table 1, but there is no discussion of these estimates as a result of the questions that surround their robustness. EMPIRICAL

RESULTS

Table 2 lists the coefficient estimates from the wage equation for the entire sample as well as for the union and nonunion sectors separately. ConTable 1. Union-nonunion

Subgroup ALL MALES FEMALES DIVISION 1 DIVISION 2 DIVISION 3 PRIVATE PUBLIC FULL PROFESSOR ASSOC PROFESSOR ASSIST PROFESSOR INSTRUCTOR TENURED NONTENURED BLACK WHITE EAST NORTHCENTRAL WEST SOUTH ECONOMICS

Sample 2988 2525 463 2283 640 65 702 2286 1179 870 797 122 2117 871 51 2937 840 963 391 794 150

percentage

size

salary

spicuous was the large and significant coefficient for the ADVANCED DEGREE variable in the nonunion sector compared with the smaller and statistically insignificant coefficient for that variable in the union sample. YRSINRANK was highly significant as expected in both sectors as were the three faculty ranks of Assistant, Associate and Full Professor (Instructor was the omitted base variable). Although TENURED was negative across both sectors, the coefficient was insignificant signifying that tenure per se has little impact on faculty salaries in either sector. In both sectors, the existence of a heavy teaching load led to lower salaries, though more so in the nonunion sample. This provides evidence that the market-place provides higher rewards to faculty who are research-oriented, since it is well-known that heavier teaching loads are characteristic of institutions where research is not a major focus. It was expected that the RESEARCH variable, which identified faculty as primarily teaching-oriented or research-oriented would reinforce this finding. The coefficient estimate turned out, however, to be insignificant in both samples. The number of articles and books published were differentials

Percentage salary Model I* 4.40 4.32 3.28 3.70 11.51 -1.88 13.07 4.09 8.94 4.15 2.10 -3.08 5.11 0.36 11.91 4.45 5.87 -1.19 -7.63 7.92 0.51

(4.25) (3.70) (1.55) (3.07) (5.17) (-0.28) (2.46) (4.16) (5.27) (2.44) (1.04) (-0.49) (3.94) (0.15) (1.32) (4.23) (3.03) (-0.64) (-1.00) (2.72) (0.13)

for different differential Model II* 4.18 8.65 1.51 3.66 0.17 3.03 13.00 5.34 2.52 7.09 4.56 4.18 7.62 -2.43 -3.37 8.02 -

faculty

subgroups Percentage of faculty from an organized campus 14.03 13.24 18.10 12.13 13.62 75.26 3.29 17.24 11.93 16.76 13.68 17.21 14.93 11.88 16.67 13.97 1.21 10.41 4.99 4.93 12.35

*Model I and Model II refer to conventional and simultaneous equations estimates, respectively. Note: Figures in parentheses are t-statistics corresponding to the union coefficient from which the union-nonunion differentials were calculated. Some of the Model II estimates could not be calculated due to the small number observations in some cells.

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Do Unions Affect Faculty Salaries? Table 2. Wage equations for the union, nonunion and entire sample

Variable Intercept Advanced degree Experience Exp. squared Years in rank Assistant Professor Associate Professor Full Professor Tenured Heavy load Books Articles Male Age Research Private Division 1 Division 2 Black Union R2 No.

Union sample Estimate r-statistic

Nonunion sample t-statistic Estimate

7.1010 0.0197 0.0077 -0.0002 0.0094 0.0730 0.2429 0.4369 -0.0081 -0.0224 0.0052 0.0028 0.0469 0.0051 0.0149 0.0211 -0.0977 -0.1099 0.0212

7.3963 0.0549 0.0019 -0.0001 0.0069 0.0635 0.2178 0.4062 -0.0189 -0.0470 0.0104 0.0030 0.0601 -0.0024 -0.0053 0.0172 -0.2489 -0.3123 0.0292

of observations

114.19* 0.91 2.04** -2.67* 4.55* 1.91*** 5.62* 9.21* -0.32 -1.26 1.56 3.62* 2.32** 4.13* 0.77 0.54 -3.49* -3.58” 0.37 0.722 405

133.82* 4.83* 1.03 -2.45** 6.50* 3.09* 9.42* 16.17* -1.48 -5.66* 6.46* 9.75* 5.54* 3.39* -0.61 2.10** -5.16* -6.44* 1.03 0.619 2583

Entire sample Estimate r-statistic 7.2660 0.0517 0.0028 -0.0001 0.0072 0.0628 0.2166 0.4055 -0.0171 -0.0435 0.0094 0.0029 0.0568 0.0029 -0.0027 0.0190 -0.1396 -0.1954 0.0252 0.0431

202.42* 5.07* 1.69*** -3.34* 7.62* 3.43* 10.52* 18.06* -1.48 -5.74* 6.44* 10.23* 5.86* 4.65* -0.35 2.41** -5.73” -7.91* 0.98 4.25*

0.626 2988

*Significant at the 0.01 level. **Significant at the 0.05 level. ***Significant at the 0.10 level.

both very significant in explaining higher salaries across both sectors. This result was much stronger, however, in the nonunion sector. The difference between salaries at private institutions relative to public ones was statistically insignificant in the union sector. However, the coefficient for private institutions was positive and significant in the nonunion sector. In other words, all other things equal, private schools have higher salaries than public institutions only in the nonunion sector. Interestingly, while 17.24% of all public school faculty were from organized schools, only 3.29% of private schools were unionized. There was little evidence of discrimination against blacks in either sector. Being male, however, had a significant positive influence on salaries in both sectors, with the effect being stronger in the nonunion sector. The negative and significant coefficients for DIVl and DIV2 in both sectors suggest that DIV3 faculty were compensated better than their counterparts in the other two divisions, holding constant their credentials and characteristics.

It was clear that faculty characteristics had a much larger impact on salaries in the nonunion relative to the union sector. In particular, HVYLOAD, BOOKS and PRIVATE were highly significant in the nonunion sector, but insignificant in the union sample. Most of the other variables also had larger absolute values and higher levels of significance in the nonunion sample. Union-Nonunion

Salary Differentials

Estimates from both the single-equation OLS model and the simultaneous equations model are reported in Table 1. As mentioned previously, the latter results (Model II) have not been discussed due to concerns about their robustness. Separate regressions were run for each subgroup listed in the table. The union-nonunion salary differential of 4.40% for the entire sample represents a broad average. As Table 1 indicates, the union wage premium varied from a low of -7.63% for schools in the west to a high of 13.07% for private schools. A prominent result was that unions extracted an

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11.51% premium for faculty in Division 2 schools. The union salary mark-up was only 3.70% in Division 1 institutions, and the difference was statistically insignificant in Division 3 schools. These results are not surprising. Among Division 3 schools, 75.26% of the respondents reported association with a unionized school. The “threat effect” has been known to induce nonunion employers to keep compensation of their workers comparable with levels of union employees. Since, ceteris paribus, DIV3 faculty already had the highest salaries (discussed earlier in reference to Table 2), it appears that unions were not able to raise their salaries any further. The pay-off to being in a unionized institution increased with rank. The gain to Instructors from collective bargaining was statistically insignificant. It rose to 2.10% for Assistant Professors, 4.15% for Associate Professors and climbed to 8.94% for Full Professors. This is not consistent with the results of many earlier studies on the wage effects of unions. Freeman (1980), for example, found that unions have a propensity to standardize wages and thus reduce wage dispersion. Groups that have traditionally been lower on the ladder of economic success (blacks, women, blue-collar workers, etc.) stand to gain more than their more privileged workmates who have not been subjected to labor market discrimination. This follows from the fact that in the setting of a discrimination-free union, they have more “catching-up” to do. For these reasons, one would have expected that the lower the rank, the greater the gain the faculty member would derive from unionism. Nonetheless, while unions may not have affected faculty across different ranks in the manner they have in other occupations, this article does support another study by Freeman (1978) whose results across faculty ranks were similar to those in this article. It also supports findings by Barbezat (1989). Tenured faculty stood to gain more from unionism than did their nontenured counterparts (5.11% for tenured as opposed to 0.36% for nontenured). This runs counter to the Freeman argument of a reduction in wage dispersion as a result of unions, since the earnings differentials between tenured faculty at unionized and non-unionized schools is greater than the gap for untenured faculty. Freeman’s argument, however, was in regard to

unions in general, and has been supported by several studies. The results here indicate that faculty unions are an exception to the rule. Black faculty had a large 11.91% gain from association with a unionized institution as opposed to 4.45% for Whites (the small number of observations for black faculty, however, suggests caution in the interpretation of results). Geographical stratification of schools suggested that while unions elicited moderate wage gains in the East and the South (5.87 and 7.92%) respectively), they were much less effective in the northcentral areas and the west (-1.19 and -7.63%, respectively). Finally, as an interesting sidelight, union wage premiums were calculated for a sample of economics faculty alone. The resulting union wage premium of 0.5% (statistically insignificant) suggests that unions have had little impact on the salaries of practitioners of our specialty in the profession.

CONCLUSIONS This paper calculated union wage premiums for numerous groupings of college faculty in the United States and found the union wage premium to be a moderate and statistically significant 4.40%. This is very close to the 4% differential Freeman (1978) found for his sample of college faculty in 1976-1977, but higher than the results generally reported by Barbezat (1989). Across subgroups, the differential ranged from -7.63 to 13.07%. Faculty characteristics were found to be generally less important in influencing salaries in the union sector than among unorganized campuses. Rewards from unionism appear to increase with rank and are higher for tenured faculty a result inconsistent with the Freeman thesis that unions reduce wage dispersion. Males gain somewhat more than females from collective bargaining, while association with unionized schools is far more advantageous for blacks than it is for whites. Acknowledgement - The author wishes to express his gratitude to Professor Stephen Woodbury for extremely helpful comments. The author also acknowledges valuable advice from an anonymous referee of this journal. An earlier version of this paper was presented at the Allied Social Science Associations meetings in Washington, DC (27-30 December 1990).

Do Unions Affect

Faculty

Salaries?

NOTES 1. A confidential list of school names was provided by Professor Ladd at the University of Connecticut who was one of the principal investigators of the survey. These names were matched up with institutions that had collective bargaining contracts in 1976 according to the 1984 Directory of Faculty Contracts to determine the union status variable for institutions. 2. The survey asked the question, “Do your interests lie primarily in research or teaching?“. Responses were (1) very heavily in research; (2) in both, but leaning toward research; (3) in both, but leaning toward teaching; and (4) very heavily in teaching. For this study, responses of (1) and (2) were treated as “research-oriented” and (3) and (4) as “teaching-oriented”. 3. The AAUP defines Category 1 as doctoral-level institutions which grant a minimum of 30 doctorallevel degrees annually in three or more unrelated disciplines. Category IIA are comprehensive universities characterized by diverse post-baccalaureate programs, but that do not engage in significant doctoral-level education. Category IIB are institutions with primary emphasis on general undergraduate baccalaureate-level education (this study treats Categories IIA and IIB as one group). Category III are two-year institutions with academic rank, while Category IV are mainly 2-year colleges that do not utilize academic ranks. This study treated Categories III and IV as one group. This information was taken from Academe (Bulletin of the AAUP).

REFERENCES BARBEZAT, D. (1989) The effect of collective bargaining on salaries in higher education. Zndust. Labor Relat. 42, 443-455. DOUGLAS, J.M. and ROSENBERG, C. (1984) Directory of Faculty Contracts and Bargaining Agents in Institutions of Higher Education. New York: The National Center for the Studv of Collective Bargaining in HigTher Education, Baruch College. FREEMAN, R.B. (1978) Should we organize? The effects of faculty unionism on academic compensation. Working Paper No. 301, National Bureau of Economic Research, November. FREEMAN,~R.B. (1980) Unionism and the dispersion of wages. Indust. Labor Relat. Rev. 34(l), 3-23. FREEMAN, R.B. and MEDOFF, J.L. (1981) The impact of collective bargaining: illusion or reality? In U.S. Industrial Relations 1950-1980: A Critical Assessment (Edited by STEIBER, J., MCKENSIE, R. and MILLS, D.), pp. 47-97. Madison, WI: Industrial Relations Research Association. KELLEY, E.P. (1977) Special Report No. 12. Washington, DC: Academic Collective Bargaining Information Service. LADD, E.C. and LIPSET, S.M. (1977) 1977 Survey of the American Professoriate. Documentation for computer tape. Storrs, Connecticut: Roper Center, University of Connecticut.

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