the effect of unionization on faculty salaries and compensation: estimates from the 1980s

The Effect of Unionization on Faculty Salaries and Compensation: Estimates from the 1980s*

DANIEL I. REES University of Colorado at Denver, CO 80217

Institutional data obtained from the AAUP and other sources are used to explore the effect of unionization on faculty salaries and compensation. This research differs from previous work in that it investigates the effect of affiliation with a particular national union, estimates the impact of decertification, and employs data from the 1980s, whereas previous research primarily used data from the 1970s. The cross- section results for the mid-1980s indicate a union-nonunion compensation differential of about 5 to 6 percent. The fixed-effects results indicate a negative union-nonunion compensation differential.

I. Introduction

A number of studies have investigated the impact of unionization on faculty salaries or compensation at American colleges and universities using data from the 1970s. More recent estimates of faculty union-nonunion salary or compensation differentials do not exist, at least in published form. 1

Using data provided by the American Association of University Professors (AAUP) and other sources, I estimate union-nonunion salary and compensation differentials for academic years 1985-1986, 1986-1987, and 1987-1988. A fixed-effects model is also estimated to obtain the average differential for a longer period, 1970- 1971 through 1987-1988.

II. Background

Faculty unionism in U.S. higher education first occurred in 1963 when the faculty of Milwaukee Area Technical College elected the American Federation of Teachers (AFT) as their bargaining agent. Faculty at other institutions soon followed; by the mid-1970s, over 300 colleges and universities were organized (Douglas, 1990, p. 111). This rapid growth of faculty unionism inspired a large body of research investigating the effect of unionization on faculty salaries and benefits. Interestingly, no clear pattern emerged. Birnbaum (1974, 1976), Morgan and Kearny (1977), and Leslie and Hu (1977) found positive union effects on faculty salary or compensation. But Marshall (1979), Guthrie-Morse et al. (1981), Hu and Leslie (1982), and Kesselring (1991) found either no evidence of a union effect or a negative effect.

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These studies suffered from methodological problems: the sample sizes were typically quite small, and, except for Kesselring (1991), all of the studies relied, at least in part, on the matching procedure, which has the drawback of forcing the researcher to make subjective judgements about the pairing of institutions. 2

Barbezat (1989) and Freeman (1978) employed larger samples and avoided the matching procedure. Barbezat's work is unique in that it used individual-level data, which allowed her to control for faculty characteristics, such as the number of articles published, sex, race, and education. Her estimates suggested that unionization had little effect on the level of faculty salaries in 1977. 3

Freeman used information from the AAUP on approximately 900 institutions to estimate union-nonunion salary and compensation differentials by year and faculty rank, and reported positive and statistically significant differentials for each year, 1970-1971 through 1976-1977, for all faculty ranks. However, Freeman noted that the union effect seemed to grow weaker over the period examined. For instance, in academic year 1970-1971 the estimated compensation differential across all ranks was 11 percent, while in 1976-1977 it was only 5 percent.

Freeman also estimated an average compensation differential for the period 1970- 1971 through 1976-1977 using a fixed-effects approach. The fixed-effects results confirmed that unionization increased compensation, but the magnitude of this effect was substantially smaller than that obtained from data on specific academic years.

III. Data

The primary data source for this essay is the AAUP's Annual Survey of Faculty Compensation. 4 Each year the AAUP asks colleges and universities across the U.S. to report, by rank, the number of full-time faculty they employ and the mean salary and compensation levels of their full-time faculty.

Information on when a school's faculty voted to be represented by a bargaining agent comes from the Directory of Faculty Contracts (Douglas, 1990), which lists all collective bargaining agreements between faculty and institutions of higher education in the U.S., and all instances of a faculty vote in favor of unionization. (Other data sources are given in the appendix.)

IV. Cross-Section Model

Following Freeman (1978), a standard wage or salary equation is estimated using OLS:

lnW s = et + b(Unions) +/3'X s + es" (1)

The subscript s denotes school, and W is equal to either the mean faculty salary at the school or the mean of faculty compensation. (Faculty compensation is salary plus the cost of faculty benefits. 5) In the equations estimated separately by rank, salary and compensation means for faculty in the relevant rank are used.

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Union is equal to 1 for schools whose faculty were represented by a bargaining agent and equal to 0 otherwise. I f a school's faculty were represented by a bargaining agent, but had not yet signed a formal collective bargaining agreement, Union was nevertheless set equal to 1. The estimated coefficient of Union will be approximately equal to the average salary or compensation union-nonunion differential in percentage terms. In some estimations the variables Years Organized and Years Organized Squared are added to equation (1) in order to explore how the impact of unionization is affected by the length of time a school has been organized.

X s is a vector of controls which are similar, but not identical, to those used by Freeman (1978). More specifically:

Faculty Size is the number of full-time faculty at the school and is included in order to control for institutional size. Numerous studies have documented a positive relationship between establishment size and the salary or wage paid to employees (see, for instance, Mellow, 1982; Brown and Medoff, 1989).

Fraction of Faculty with Ph.D.s is a proxy for the quality of a school's faculty. Presumably, schools seeking high-quality faculty will pay relatively high salaries or benefits.

Faculty-Student Ratio controls for the average teaching workload. 6 The theory of compensating differentials suggests that lighter workloads should be associated with lower pay. However, teaching workloads may also be related to institutional quality, and so the relationship between the faculty-student ratio and faculty compensation may not be negative.

Also included are variables measuring the type of degree granted by the institution, undergraduate admissions selectivity, and whether the institution was privately controlled or affiliated with a religious organization. These factors can potentially influence the type of labor market in which a school operates. Variables for region (Northeast, Northcentral, and West), and whether the campus was situated in a large city or small town/rural setting are used in order to control for cost of liv- ing differences.

In the equations estimated separately by rank, controls are included for the proportion of faculty members with tenure and the proportion of female faculty. In the equations estimated across all ranks, controls for the proportion of faculty who were full professors, associate professors, and assistant professors are used. Also, a dummy variable equal to 1 if the school had no faculty ranks, and equal to 0 otherwise, is included. (The variables are defined in the appendix.)

V. Empirical Results: Cross-Section Analysis

Means and standard deviations of selected variables are shown in Table 1 by the union status of the school's faculty. The average salary and compensation levels at unionized schools are somewhat higher than at their nonunion counterparts. How-


Table 1

Means and Standard Deviations o f Selected Variables in the 1987-1988

Cross-Section Analysis, All Ranks, by Union Status

Unionized Schools Nonunion Schools (n = 346) (n = 1,250)

Standard Standard Variable Mean Deviation Mean Deviation

Union 1 0 0 0

Years Organized 13.0 4.6 0 0

Salary 33,610 6,149 29,731 7,014

Compensation 41,436 7,886 36,176 8,951

Faculty Size 218 271 182 280

Faculty-Student Ratio .05 .02 .10 .96

Fraction Faculty w/Ph.D. .32 .30 .45 .31

Campus in Large City .14 .34 .17 .38

Campus in Small Town .22 .42 .22 .42 or Rural Setting

Category E (2-year) .54 .50 .25 .15

Privately Controlled .07 .26 .25 .43

Church Affiliated .02 .15 .31 .46

AAUP .13 .34 0 0

AFT .30 .46 0 0

NEA .44 .50 0 0

Other .06 .23 0 0

Joint .07 .25 0 0

ever, there are other differences between these two types of schools that must be taken into account before these differences can be at t r ibuted to unionism. For instance, unionized schools are, on average, larger than nonunion schools, are less likely to be privately controlled, and are more likely to offer only a two-year degree.

Table 2 presents faculty compensation equations estimated separately by rank for the year 1987-1988. The estimated coefficient of the union dummy is .053 in the full professor equation which suggests that, after controlling for factors such as faculty size, institutional control, and type of degree granted, unionization increases the mean compensation level of full professors by 5.3 percent. The corresponding figures for associate and assistant professors are 6.2 and 4.2 percent, respectively. 7

D A N I E L I. REES 403

T a b l e 2

The D e t e r m i n a n t s o f F a c u l t y Compensa t ion , 1 9 8 7 - 1 9 8 8

( a b s o l u t e v a l u e t - s t a t i s t i c s in p a r e n t h e s e s )

Full Associate Assistant All Variable Professors Professors Professors Ranks

Intercept 6.155 6.023 6.016 5.515 (97.7) (193.8) (231.4) (107.1)

Union .053 .062 .042 .060 (5.3) (6.6) (4.2) (6.5)

Faculty Size a .004 .001 -.006 -.004 (4.4) (1.5) (7.4) (4.5)

Fraction Total .026 .030 .044 .042 Faculty w/Ph.D.s (2.2) (2.5) (3.5) (3.4)

Fraction of Rank -.211 -. 183 -.205 - - Female (4.3) (5.3) (6.4)

Fraction of Rank .148 .096 .041 - - Tenured (2.2) (3.9) (1.8)

Tenure Information - . 152 -. 175 -.058 - - Not Available (2.2) (6.1) (4.9)

Faculty-Student Ratio -.012 -.007 .002 -.007 (2.5) (1.1) (0.3) (1.3)

Privately Controlled .015 -.008 -.062 -.043 (1.1) (0.7) (4.9) (3.6)

Church Affiliated -.062 -.067 -. 112 -. 107 (4.0) (5.2) (8.7) (8.0)

Campus in Large City .033 .030 .025 .023 (3.9) (3.8) (2.9) (2.9)

Campus in Small -.037 -.022 -.028 -.042 Town or Rural Setting (2.8) (2.0) (2.5) (3.8)

Region Northeast .076 .072 .066 .075 (6.7) (6.9) (6.1) (7.0)

Region Northcentral -.006 .003 .013 .018 (0.6) (0.3) (1.5) (2.1)

Region West .056 .037 .055 .049 (5.2) (3.4) (4.6) (4.4)

Admissions I .260 .194 .182 .260 (9.6) (6.5) (6.2) (9.4)

Admissions 2 .181 .118 .131 A80 (7.9) (5.1) (5.4) (8.1)

Admissions 3 .094 .067 .074 .107 (4.6) (3.3) (3.4) (5.5)

Admissions 4 .016 .007 .005 .022 (0.8) (0.4) (0.2) (1.3)

4 0 4 J O U R N A L O F L A B O R R E S E A R C H

Table 2 - - Continued

Admissions 5 - .060 - .043 - .059 - .045

(2.9) (2.2) (2.8) (2.3)

Admissions 6 - .099 - .059 - .067 - .063 (3.9) (2.5) (2.7) (2.6)

Category B - .085 - .053 - . 105 - . 109

(8.0) (5.5) (10.3) (11.3)

Category C - . 186 - . 158 - . 193 - .204

(12.1) (11.8) (14.2) (15.1)

Category E - . 136 - .096 - . 142 - . 161 (4.9) (4.0) (5.4) (7.0)

n 1247 1262 1263 1596

Adjusted R-Squared .715 .557 .528 .699

Notes: The dependent variable is the natural log of mean total compensation for the relevant rank. All observations are weighted by the number of fun-time professors in the relevant rank. See the appendix for sources and definitions of variables.

The All Ranks equation also includes controls for the fraction of faculty belonging to the rank of full professor, associate professor, and assistant professor, and a dummy variable equal to 1 for schools with no faculty ranks and zero otherwise.

acoefficient multiplied by 100.

Further noteworthy results from Table 2 are:

• Faculty size is positively related to compensation for full and associate professors. For assistant professors, increases in faculty size seem to be associated with lower compensation levels.

• As expected, faculty quality, as measured by the percent of faculty with a Ph.D., is positively related to compensation.

• The larger the proportion of female professors at a school, the less the school tends to pay its faculty. 8

• The larger the proportion of tenured professors at a school, the more the school tends to pay its faculty.

• The faculty-student ratio is negatively related to the compensation of full professors. The relationships between associate and assistant professor compensation and the faculty-student ratio are not statistically significant.

• Schools with the most selective undergraduate admissions tend to pay their faculty more.

DANIEL I. REES 405

• Doctoral grant ing universi t ies tend to have the highest levels o f compensa t ion .

These resul ts were obta ined us ing m e a n total compensa t ion as the dependen t var iable . W h e n m e a n salary is used ins tead (Table 3), a s imi lar pat tern of resul ts emerges : for e x a m p l e , there is l i t t le d i f f e r ence b e t w e e n u n i o n i z a t i o n ' s ef fect o n salary and its effect on compensa t ion , at least for the year 1987-1988.

In c o l u m n (4) of Tables 2 and 3 c o m p e n s a t i o n and salary equa t ions are est i- m a t e d across all f acu l ty ranks . I n c l u d e d in these e s t i m a t i o n s are f acu l ty at 313 schools wi thout facul ty ranks, and facul ty wi th the r ank o f instructor. The overa l l

Table 3

The D e t e r m i n a n t s o f F a c u l t y Sa lar i e s : 1 9 8 7 - 1 9 8 8

(absolute value t-statistics in parentheses)

Full Associate Assistant All VariabIe Professors Professors Professors Ranks

Intercept 5.999 5.850 5.842 5.364 (96.6) (208.8) (254.9) (112.8)

Union .054 .058 .037 .059 (5.9) (6.9) (4.2) (7.0)

Faculty Size a .004 -.002 -.007 -.004 (4.6) (2.1) (9.6) (5.2)

Fraction Total .015 .017 .032 .032 Faculty w/Ph.D.s (1.3) (1.6) (2.9) (2.8)

Fraction of Rank -.163 -.172 -.218 - - Female (3.6) (5.5) (7.7)

Fraction of Rank .120 .078 .016 - - Tenured (2.0) (3.5) (0.7)

Tenure Information -. 159 -. 169 -.066 - - Not Available (2.5) (6.6) (6.4)

Faculty-Student Ratio -.010 -.006 -.000 -.007 (2.3) (0.9) (0.0) (1.4)

Privately Controlled -.003 -.018 -.056 -.050 (0,2) (1.7) (5.0) (4.4)

Church Affiliated -.071 -.067 -.094 -. 103 (5.0) (5.7) (8.2) (8.3)

Campus in Large City .037 .029 .023 .027 (4.7) (4.1) (3.0) (3.7)

Campus in Small -.045 -.028 -.034 -.047 Town or Rural Setting (3.7) (2.8) (3.4) (4.7)


Table 3 - - Cont inued

Region Northeast .064 .060 .055 .063 (6.0) (6.4) (5.6) (6.4)

Region Northcentral - .002 .008 .017 .023 (0.2) (1.0) (2.2) (3.0)

Region West .056 .027 .038 .045 (5.6) (2.7) (3.7) (4.5)

Admissions I .282 .198 .181 .270 (11.3) (7.3) (7.0) (10.5)

Admissions 2 .195 .123 .141 .187 (9.2) (5.9) (6.6) (9.1)

Admissions 3 .095 .063 .070 .103 (5.0) (3.5) (3.7) (5.7)

Admissions 4 .014 .007 .005 .021 (0.7) (0.4) (0.3) (1.2)

Admissions 5 -.057 -.043 -.056 -.046 (2.9) (2.5) (3.1) (2.6)

Admissions 6 -.091 -.045 -.049 -.053 (3.8) (2.1) (2.2) (2.4)

Category B -.097 -.060 -. 112 -. 117 (9.9) (6.9) (12.4) (13.2)

Category C -. 194 -. 157 -. 193 -.206 (13.6) (13.1) (16.2) (16.6)

Category E -.174 -.124 -.169 -.194 (6.8) (5.7) (7.2) (9.1)

n 1247 1262 1263 1596 Adjusted R-Squared .715 .583 .517 .728

acoefficient multiplied by 100.

es t imated union ef fec t on compensa t ion is 6.0 percent. The cor responding f igure for

salary is 5.9 percent.

For all facul ty ranks combined , the analysis is ex tended to the years 1985-1986

and 1986-1987 in Table 4. Resul t s s imi lar to those o f 1987-1988 are found, al though

the re is e v i d e n c e tha t u n i o n i z a t i o n ' s e f f ec t on to ta l c o m p e n s a t i o n is s o m e w h a t

greater than its e f fec t on salary in these earl ier years.

Wha t is str iking about the es t imates in Table 4 is that they are so s imilar to those

ob ta ined by F r e e m a n (1978) for the mid-1970s . A l t h o u g h F r e e m a n es t imated an

DANIEL I. REES 407

Table 4

Cross-Section Estimates of the Effect of Unionization on Faculty Compensation and Salary, All Ranks, 1985-1986 through 1987-1988

(absolute t-statistics in parentheses)

Sample Size Coefficient of Union

Year All Schools Union Schools Compensation Salary

1985-1986 1727 387 .049 (6.1) .043 (5.6)

1986-1987 1440 328 .050 (5.9) .046 (5.8)

1987-1988 1596 346 .060 (6.5) .059 (7.0)

Notes: The controls used are the same as those in Table 2 for All Ranks. All observations are weighted by the number of full-time professors in the institution. Sources are given in the appendix.

overall faculty compensation differential as high as 11 percent, this estimate was for the first year he examined, 1970-1971. In later years the differential seemed to shrink, until by 1976-1977 it had fallen to 5 percent, the same estimate as I obtained for the academic years 1985-1986 and 1986-1987.

The market for university and college faculty was severely depressed throughout the 1970s; average real salaries fell by more than 15 percent from 1970-1971 to 1981-1982 (Hansen, 1986). Beginning in the early 1980s, however, the academic market improved, and real salaries advanced steadily through the mid-1980s. 9 Given these contrasting economic environments, it is surprising how little faculty union bargaining power seems to have changed over the period. 1°

VI. The Effect of Years Organized on the Union-Nonunion Differential

Freeman (1978, pp. 17-19) found that newly-formed unions were less successful than their older counterparts in raising members' salary and compensation levels. An extra year of being organized was associated with an increase in faculty compensation of 2 to 3 percent. However, he raised the possibility that the advantages of growing older might not go on indefinitely.

The mean length of time organized for the union schools in my sample is between 11 and 13 years, depending on the year examined. In Freeman's sample the mean number of years organized was much shorter - - between .75 and 3.7 years. If at some point faculty unions face diminishing returns to growing a year older, then one would expect that the average increase in compensation associated with an extra year of being organized should be less in the mid-1980s than in the 1970s.

Results presented in Table 5 offer some support for the diminishing returns hypothesis. In the compensation equations without the variable Years Organized


Table 5

Cross-Section Estimates of the Effect of Years Organized on Faculty Compensation, All Ranks, 1985-1986 through 1987-1988


Est imated Coefficients M e a n Years Organ ized Years

A c a d e m i c Years Organ ized Year All Schools Union Schoo l s Un ion Organ ized Squared

t 9 8 5 - 1 9 8 6 2 .52 1 t . 26 - . 0 1 6 .006 - -

(0.9) (4.2)

1985-1986 - - - - .051 - . 0 0 9 .001

(1.7) (1.6) (2.7)

1986-1987 2.67 11.72 - . 0 1 5 .006 - -

(0.8) (3.8) 1986-1987 - - - - .122 - . 0 2 2 .001

(3.6) (3.7) (4.8)

1987-1988 2.83 13.04 - . 0 4 3 .008 - -

(2.0) (5.3)

1987-1988 - - - - .216 - . 0 4 0 .002

(5.5) (6.3) (7.8)

Notes: The controls used are the same as those in Table 2 for All Ranks. The dependent variable is the natural log of mean compensation. All observations are weighted by the number of full-time professors in the institution. Sample sizes are given in Table 4, and sources are given in the appendix.

Squared, the estimated coefficient of Years Organized is between .006 and .008, suggesting an average return to an extra year of being organized of less than 1 percent. However, when Years Organized Squared is added, the results indicate an initial rise in compensation after a representation election, followed by a period in which compensation drops, then, 6 to 8 years after an election, a period of steadily rising compensation. 11

In order to confirm this pattern seven dummies were added to the equation, each corresponding to a three-year interval (Table 6). For example, if a school had been organized for 1 year, the dummy covering the 0 to 3 year range would equal 1 and the other dummies would equal 0. Again, the results indicate an initial increase in compensation, followed by a drop almost to the level before unionization, and then an increase. This last increase is substantial: for instance, the results for 1987-1988 indicate a 21 percent salary advantage for schools organized more than 18 years.

Why the relationship between years organized and compensation follows this peculiar path is unclear. Perhaps it is a reflection of unmeasured factors affecting all schools that organized in particular years.


VII. Cross-Section Estimates of the Union Effect by Institution Control and Type

The analysis has thus far grouped all schools together, but faculty union power might differ according to school type. For instance, it has been suggested that public sector unions should be more successful than their private sector counterparts in extracting wage premiums because "the market does not discipline the union in the public sector to the extent that it does in the private" (Wellington and Winter, 1969, p. 1123).

Table 7 provides estimates of compensation premiums according to whether the institution was publicly or privately controlled, and whether it offered four- or two-year degrees. The results show that there is virtually no difference between the union effect in the private and public sectors. In each sector the union-nonunion compensation differential is approximately 6 percent, and the return to an extra year of being organized is a little less than 1 percent. 12

Table 6

Cross-Section Estimates of the Effect of Years Organized on Faculty Compensation, All Ranks, 1985-1986 through 1987-1988

(absolute t-statistics in parenthesis)

Years Organized 1985-1986 1986-1987 1987-1988

0 to 3 -.009 .023 .005 (0.2) (0.5) (0.1)

3 to 6 .044 .080 .117 (2.2) (3.7) (5.3)

6 to 9 .057 .008 -.015 (3.5) (0.4) (0.5)

9 to 12 .021 .022 .017 (1.8) (1.9) (1.2)

12 to 15 .051 .060 .028 (3.8) (3.7) (1.6)

15 to 18 .105 .083 .070 (5.7) (5.5) (5.0)

18 and up .173 .145 .207 (5.4) (5.4) (10.3)

n 1,727 1,440 1,596 Adjusted R-Squared .683 .706 .715

Notes: Controls are the same as in Table 2 for All Ranks. The dependent variable is the natural log of mean compensation. All observations are weighted by the number of full-time professors in the institution. Sources are given in the appendix.

410 J O U R N A L OF L A B O R RESEARCH

Table 7

Cross-Section Estimates of the Impact of Unionism and Years Organized on Faculty Compensation, All Ranks, 1987-1988

(Sample Divided by Institution Type and Control)

Estimated Absolute Value Coefficient t-statistic

Publicly Controlled Schools (n = 863)

Union (w/o Years Organized) .062 5.2

Union -.047 1.8

Years Organized .009 4.7

Private and Church Affiliated Schools (n = 733)


Union -.055 -0.7


Two-Year Schools (n = 497)


Union -.097 2.8


Four-Year Schools (n = 1,099)


Union .024 0.8


Notes: All observations are weighted by the number of full-time professors in the institution. The dependent variable is the natural log of mean compensation, Controls are the same as those used in Table 2 for All Ranks. Sources are given in the appendix.

There is, however, a difference between two- and four-year schools. The union- nonunion compensation differential for two-year schools is about 7.4 percent, and the return associated with an extra year of being organized is 1.3 percent. In four-year schools the differential is 3.6 percentage points smaller, and newly organized unions are almost as effective in raising faculty compensation as unions that have been in existence for a number of years. 13

VIII . Does Union Affiliation Matter?

Most faculty bargaining units are affiliated with one of three national unions: the American Association of University Professors (AAUP), the American Federation of Teachers (AFT), or the National Education Association (NEA). About 10 percent of faculty bargaining units are independent or affiliated with a national union other than

DANIEL I. REES 411

the AAUP, AFT, or NEA, and less than 2 percent have joint affiliation with two of the three primary national unions. 14

Although some research has been done on the types of clauses that are typically found in each national union's contracts (Bognanno et al., 1978, p.199), there has been no empirical investigation of whether affiliation with a particular national union has an impact on salary or compensation. In order to conduct such an investigation, in Table 8 the union dummy is replaced by 5 dummies, each corresponding to a different affiliation. 15

The pattern of results in Table 8 suggests that faculty unions that are jointly affiliated with two of the three primary organizations do the best: on average they enjoy a compensation premium of 13.4 percent and a salary premium of 14.5 percent. This may be because affiliation with more than one national union provides extra support. However, there also exists the possibility that the national unions are more willing to share representation of strong bargaining units.

Affiliation with either the AFT or NEA is associated with approximately the same increase in salary and compensation levels, e.g., the estimated coefficients of AFT and NEA are .060 and .053, respectively, in the compensation equation. Affilia- tion with the AAUP is associated with somewhat lower levels of salary and compensation. Faculty belonging to independent unions or employee organizations other than the AAUP, AFT, and NEA seem to have compensation and salary levels that are, on average, the same as faculty at nonunion schools. 16

Table 8

Cross-Section Estimates of the Effect Bargaining Unit Affiliation on Faculty Compensation and Salary, All Ranks, 1987-1988


Estimated Coefficient Bargaining Agent Salary Compensation

AAUP .039 (2.8) .047 (3.2)

AFT .062 (4.9) .060 (4.4)

NEA .052 (4.3) .053 (4.1)

Other -.026 (0.7) -.017 (0.4)

Joint .145 (7.2) .134 (6.1)

n 1596 1596

Adjusted R-squared .732 .702

Notes: Controls are the same as in Table 2 for All Ranks. All observations are weighted by the number of full-time professors in the institution. Sources are given in the appendix.

412 J O U R N A L OF L A B O R R E S E A R C H

IX. The Fixed-Effects Model

The cross-section results reported in Tables 2 through 8 strongly suggest the existence of a faculty union compensation differential of 4 to 7 percent, depending on the year and type of school. While not as large as the 15 percent figure often cited for the typical union worker in the United States, 17 a premium of 4 to 7 percent is not negligible.

However, cross-section estimates of the union effect may be biased. Freeman (1978, p. 11) reported that, holding other factors constant, faculty compensation was positively related to the probability that a school would become unionized in the future, a finding which suggests the presence of an unobserved variable that is positively correlated with both faculty compensation and unionization. If this unobserved variable is assumed constant over time, then a new model of faculty compensation can be written:

lnWst = a + b(Unionst ) + [~'Xst + v s + Est, (2)

where v s is the unobserved variable. This "fixed-effects" model is easily estimated taking differences from the mean (see Brown (1980) for an explanation of this tech-

Table 9

Characteristics of Schools in the Fixed-Effects Sample 1970-1971 through 1987-1988

Four-Year Two Year

Number of schools 1,704 1,182 Number of changers 174 207 Number always union 66 226 Number always non-union 1,464 749

Private and Publicly Church Affiliated Controlled

Number of schools 1,376 1,478 Number of changers 48 329 Number always union 31 260 Number always non-union 1,297 889

Entire Sample

Number of schools 2,886 Number of changers 381 Number always union 292 Number always non-union 2,213

Note: A "changer" is a school that went from union to nonunion status or vice versa.

DANIEL I. REES 413

nique), which, in effect, produces an intercept for each school in the sample. These school-specific intercepts control for variables that are constant over time, including the unobserved variable, Vs .18

The fixed-effects analysis uses data from 2,886 schools. Only those schools that reported information at least twice during the period 1970-1971 through 1987-1988 are included. Table 9 gives the number of schools that changed union status according to type of degree offered and whether the school was public or private. The coefficient of the union dummy depends on the presence of these "changers" for its identification. Although there are fully 2,886 schools in the sample, only 381 of them went from nonunion to union status or vice versa. Of these 381 schools, 207 were two-year, and 329 were publicly controlled; 292 schools were unionized throughout the period of analysis.

Most of the independent variables used in the previous sections are constant over time and, therefore, cannot be used in the fixed-effects analysis. If included in the vector Xst they would be perfectly colinear with the school-specific intercepts. Controls that are not constant over time and for which information is available are: institution type, faculty size, year dummies, and the percent of faculty with the rank of full professor, associate professor, and assistant professor.

Although short, this list includes several controls that were not used in Freeman's fixed-effects analysis. Freeman used only institution type and year dummies as con- trois. Another problem is that Freeman did not weight his observations by the number of full-time faculty at the institution - - a necessary step in order to avoid heteroskedasticity. For comparison purposes, I present estimates of compensation equations that use weighted observations and the full set of available controls, as well as estimates that use unweighted observations and Freeman's more limited set of controls.

X. The Fixed-Effects Results

As shown in Table 10, my specification produces results that differ dramatically from Freeman's. Columns (1) and (2) present estimates using the Freeman specification. Here, the union effect seems to be positive: a change from nonunion to union status is associated with a statistically significant increase in compensation of 1.2 percent. When Years Organized is added to the equation, its coefficient is negative, but not significant, and the coefficient of the union dummy retains its magnitude and sign.

In contrast, the estimates from the full model (columns 3 and 4) suggest that unionization results in a decrease in compensation. Going from nonunion to union status is associated with a statistically significant drop in compensation of .5 percent. And, as the number of years a school has been organized increases, the union effect grows more negative. For example, in the first year after an election there seems to be no union effect, but after 10 years a unionized school's compensation level would have fallen by about 2 percent. 19


Table 10

Fixed-Effects Estimates of the Impact of Unionization and Years Organized on Faculty Compensation, All Ranks,

1970-1971 through 1987-1988 (absolute t-statistics in parentheses)

Variable ( l ) (2) (3) (4)

School Dummies yes yes yes yes

Year Dummies yes yes yes yes

Weighted no no yes yes by Faculty Size

Category A .031 .031 .045 .045 (4.1) (4.1) (9.9) (9.8)

Category B .012 .012 .021 .022 (4.1) (4.1) (7.0) (7.2)

Category C .002 .002 .012 .011 (0.5) (0.5) (2.3) (2.1)

Fraction Full - - - - .180 .181 (30.1) (30.3)

Fraction Associate - - - - .049 .051 (7.6) (7.9)

Fraction Assistant - - - - - .077 - .079 (18.4) (18.8)

Faculty Size a - - - - .008 .005 (1.9) (1.2)

Union .012 .012 - .005 .002 (4.8) (4.5) (3.0) (0.8)

Years Organized - - - . 0 0 0 - - - .002 (0.0) (10.7)

R-Squared .927 .927 .934 .935

DOF 30,388 30,387 30,384 30,383

n 33,295 33,295 33,295 33,295

Notes: The dependent variable is the natural log of mean compensation. Sources are given in the appendix.

acoefficient multiplied by 1000

Clearly, within the fixed-effects framework, the finding of a positive compensation differential is not particularly robust. It is difficult, however, to understand why unionization might cause a drop in compensation levels. One explanation is that faculty members decide to unionize when an unobserved event occurs that promises to decrease future compensation. Compensation and unionization could then have a negative relationship, although the relationship would not be causal. 20

DANIEL I. REES 415

The full fixed-effects model is also estimated separately by institution type and

control (Table 11). Interest ingly, un ion iza t ion is associated with an increase in compensat ion levels for private and church-affil iated schools. At public schools, two- and four-year schools, however, unionization is associated with a decrease in compensation.

Finally, when the analysis of faculty compensation was restricted to the years examined by Freeman (1970-1971 through 1976-1977), 21 Freeman 's specification still yielded an average union effect of about 1 percent. The full model produced a positive but insignificant estimated union coefficient, although when Years Organized was added to the equation its coefficient was positive and significant.

Table 11

Fixed-Effects Estimates of the Impact of Unionization and Years Organized on Faculty Compensation, All Ranks,

1970-1971 through 1987-1988 (Sample Divided by Institution Type and Control)

Estimated Absolute Value Coefficient t-statistic

Publicly Controlled Schools (n = 18,145) Union (w/o Years Organized) -.010 Union -.003 Years Organized -.003

Private and Church Affiliated Schools (n = 14,999)

4.7 1.3

11.1

Union (w/o Years Organized) .009 2.1 Union -.002 0.4 Years Organized .002 4.0

Two-Year Schools (n = 12,253) Union (w/o Years Organized) -.007 2.3 Union -.001 0.2 Years Organized -.005 15.6

Four-Year Schools (n = 21,042)

Union (w/o Years Organized) -.013 5.6 Union -.007 3.0 Years Organized -.001 5.6

Notes: All observations are weighted by the number of full-time professors in the institution. The dependent variable is the natural log of mean compensation. Controls are the same as those used in Table 10, columns (3) and (4). Sources are given in the appendix.


Thus, there is evidence of a positive return to an extra year of being organized in the earlier period. Despite this finding, however, estimates of the average union- nonunion compensation differential are quite small - - certainly much smaller than cross-section estimates of the differential.

XI. The Effect of Decertification

Most of the changers in the fixed-effects sample went from nonunion to union status, but 16 schools, all privately controlled or church-affiliated, returned to nonunion status. Their presence in the data permits the estimation of the average effect of decertification.

The sample was restricted to privately controlled and church-affiliated schools, and a dummy (Decertified = 1 in all years after a school went from union to nonunion status) was added to the basic fixed-effects model.

The results in Table 12 suggest that decertification leads to a decrease in mean faculty compensation of approximately 3.7 percent. The increase associated with organizing, however, is insubstantial (.1 percent) and statistically insignificant, so that faculty at these schools seem to be worse off than before they were unionized.

These findings might be surprising if one viewed decertification as a wholly voluntary proposition. However, the overwhelming majority of decertifications are the result of the Yeshiva 22 decision (Douglas 1990, p. 118), which withdrew the pro- tection of the National Labor Relations Act from faculty at private schools. Thus, decertification can be viewed as typically exogenous in nature. The finding that it leads to a reduction in compensation greater in magnitude than the increase associated with organizing is consistent with the hypothesis that faculty elect bargaining agents to protect them from the threat of future reductions in compensation.

XII. Conclusion

Freeman's (1978) cross-section estimates of the union-nonunion compensation differential for faculty in higher education were between 5 and 11 percent, while his fixed-effects results suggested a smaller, but nevertheless significant, union effect. I examine more recent data than Freeman and find evidence of a union-nonunion compensation differential of 6 percent in 1987-1988 and 5 percent in the years 1985-1986 and 1986-1987. These figures are similar to Freeman's cross-section estimates for the mid-1970s.

However, my fixed-effects analysis produced sharply different results from Freeman's. Adding several controls not used by Freeman, I found a negative relationship between unionization and compensation for the period 1970-1971 through 1987-1988. When the fixed-effects analysis was restricted to the period examined by Freeman (1970-1971 through 1976-1977), the addition of the new controls eliminated the finding of a positive and statistically significant union effect.

DANIEL I. REES 417

Table 12

Fixed-Effects Estimates o f the Impact o f Unionization and Decertif ication on Faculty Compensation,

Al l Ranks, 1970-1971 through 1987-1988

Private and Church-Affiliated Schools Only (absolute t-statistics in parentheses)

Variable

School Dummies yes Year Dummies yes Weighted by Faculty Size yes

Category A .009 (1.o)

Category B .016 (4.9)

Category C .030 (3.7)

Fraction Full .149 (18.3)

Fraction Associate .111 (12.0)

Fraction Assistant -.026 (3.4)

Faculty Size a .162 (12.6)

Union .001 (0.2)

Decertified -.037 (5.3)

R-Squared .958 DOF 13,597 n 14,999

Notes: The dependent variable is the natural log of mean compensation. Sources are given in the appendix.

acoefficient multiplied by 1000

This research casts doubt on the exis tence o f a posi t ive u n i o n - n o n u n i o n differential. Al though the cross-sect ion ev idence seems robust, it could be due to one or more omit ted variables. Wi thout ev idence of a longi tud ina l nature, one is left won- der ing i f the cross-sect ion controls used by myse l f and F reem an were adequate.

418 J O U R N A L OF L A B O R RESEARCH

A P P E N D I X

Variable Definitions and Sources

From the AAUP's Annual Survey of Faculty Compensation, Academic Years 1970-1971 through 1987-1988:

Faculty Size . . . . . . . . . . .

Salary . . . . . . . . . . . . . . .

CompensatiOn . . . . . . . . .

Fraction Full . . . . . . . . . .

Fraction Associate . . . . . .

Fraction Assistant . . . . . .

The total number of full-time faculty employed at the institution

The mean salary of full-time faculty, measured in 1987-1988 dollars

The mean level of compensation received by full-time faculty, measured in 1987-1988 dollars

The fraction of full-time faculty who are full professors

The fraction of full-time faculty who are associate professors.

The fraction of full-time faculty who are assistant professors.

Category B . . . . . . . . . . . . 1 if comprehensive institution, otherwise = 0 (comprehensive institutions "are characterized by diverse post-baccalaureate programs, but do not engage in significant doctoral level education")

Category C . . . . . . . . . . . . 1 if general baccalaureate institution, otherwise = 0 (these institutions concentrate on undergraduate education)

Category E . . . . . . . . . . . . 1 if two-year institution, otherwise = 0 (doctoral-level institutions comprise the omitted category)

From the AAUP's Annual Survey of Faculty Compensation, Academic Years 1985-1986 through 1987-1988:

Proportion Female . . . . . . The proportion of full-time faculty in the relevant rank who are female

Proportion Tenured . . . . . The proportion of full-time faculty in the relevant rank who are tenured (if the number of tenured faculty was not reported then this variable was set equal to 0 and a dummy variable was set equal to 1)

No Ranks . . . . . . . . . . . . . . 1 if the school has no faculty ranks, otherwise equal to 0.

Region Northeast . . . . . . . 1 if school is in northeast region, otherwise = 0

Region Northcentral . . . . . 1 if school is in northcentral region, otherwise = 0

Region West . . . . . . . . . . . 1 if school is in the west, otherwise = 0 (South is the omitted region)

Private . . . . . . . . . . . . . . . . 1 if the school is privately controlled, otherwise = 0

Church Affiliated . . . . . . . 1 if church affiliated, otherwise = 0 (public schools comprise the omitted category)

D A N I E L I. R E E S 4 1 9

From Barron 's Profiles of American Colleges (1984):

A1 . . . . . . . . . . . . . . . . . . = 1 i f most competi t ive admissions, = 0 otherwise

A2 . . . . . . . . . . . . . . . . . . . 1 i f highly competi t ive admissions, = 0 otherwise

A3 . . . . . . . . . . . . . . . . . . . 1 i f very competi t ive admissions, = 0 otherwise

A4 . . . . . . . . . . . . . . . . . . . 1 i f competi t ive admissions, = 0 otherwise

A5 . . . . . . . . . . . . . . . . . . = 1 i f less competi t ive admissions, = 0 otherwise

A6 . . . . . . . . . . . . . . . . . . = 1 i f noncompeti t ive admissions, = 0 otherwise (omitted category is

comprised of schools that special ize in music, art, or theatre, most 2-year

schools, and a few 4-year schools not included in Barron's)

From the Col lege Entrance Examinat ion Board 's Annual Survey of Colleges Research Tape, Academic Years 1985-1986 through 1988-1989:

Faculty Student R a t i o . . . The school 's overall faculty-student ratio (part-t ime students and faculty are

counted as one-half their full- t ime counterparts)

Fraction of Faculty . . . . . Proportion of full- t ime faculty who hold Ph.D.s w/Ph.D.s

Large City . . . . . . . . . . . . . 1 i f campus in ci ty with populat ion 250,000 or greater, = 0 otherwise

Small Town or Rural . . . . . 1 i f campus in town wi th populat ion 9,999 or less, = 0 otherwise

From the Directory of Faculty Contracts and Bargaining Agents in Institutions of Higher Education (1990):

Union . . . . . . . . . . . . . . . . . 1 if the school is organized, = 0 otherwise

Years Organized . . . . . . . The current year minus the year in which a school was organized (nonunion schools are assigned the value 0)

A A U P . . . . . . . . . . . . . . . . 1 i f bargaining unit was affil iated with the Amer ican Associat ion of Universi ty Professors in 1987-1988, = 0 o therwise

AFT . . . . . . . . . . . . . . . . . . 1 i f the bargaining unit was affi l iated with the American Federat ion of Teachers in 1987-1988, = 0 otherwise

NEA . . . . . . . . . . . . . . . . . 1 i f the bargaining unit was affi l iated wi th the National Educat ion Associat ion in 1987-1988, = 0 otherwise

Other . . . . . . . . . . . . . . . . . 1 i f the bargaining unit was independent or affi l iated with a national union

other than the AAUP, AFT, or N E A in 1987-1988, = 0 otherwise

Joint . . . . . . . . . . . . . . . . . . 1 if the bargaining unit was affi l iated wi th two of the three pr imary

organizations (the AAUP, AFT or NEA) in 1987-1988, = 0 otherwise

420 J O U R N A L OF L A B O R R E S E A R C H

N O T E S

*My thanks to Ronald Ehrenberg, Dennis Koepke, and Albert Rees for comments on earlier drafts. Access to the data used in this paper is limited because some school's data submissions to the AAUP are confi- dential. An appendix with additional results is available on request from the author.

lAfter writing this essay an unpublished paper by Elizabeth Davis, "A Fixed Effects Model of the Com- pensation Gains of Faculty Unions," came to my attention. Davis used data from the AAUP for the years 1967 to 1986 to estimate a fixed-effects model, but restricted her analysis to four-year schools and concen- trated on estimating the effect of signing a union contract as opposed to electing a bargaining agent, which is the focus of this as well as previous studies. Davis found that the signing of a union contract was associated with an increase in compensation of about 1 percent. Also, her results suggest that using the election of a bargaining agent as the measure of unionization would produce a union-nonunion compensation differential of less than 1 percent. Kesselring (1991) used data from 1984, but restricted his sample to institutions granting doctoral degrees.

2This same point was made by Bacharach et al. (1987, p. 249).

3One possible short-coming of Barbezat's study is that she had information on faculty from a limited number (33) of unionized institutions. These institutions may not have been fully representative of all unionized institutions.

4All studies referred to above except for Barbezat (1989) used this data source.

5Schools were asked to assess the cost of their contributions to retirement and insurance plans, tuition waivers and payments, social security taxes, unemployment taxes, workers compensation taxes, and benefits in kind with cash alternatives.

6part-time faculty and students were given one-half the weight of their full-time counterparts in the con- struction of this variable. Other weights were tried without an appreciable effect on the results.

7Barbezat (1989) found that full professors had the highest union-nonunion salary differential, followed by associate and then assistant professors. Freeman (1978) found a similar pattern of compensation differentials.

8This finding could be due to a combination of factors. For instance, it may partly reflect the fact that, on average, female faculty have entered the profession more recently than their male counterparts, or that they are more likely to teach in lower-paying disciplines, or it may be due to discrimination. A number of studies have investigated discrimination against female professors (see, for example, Gordon, Morton, and Braden, 1974).

9From 1981-1982 to 1987-1988, real faculty salaries increased by 16 percent. (Source: author's calcula- tions using the AAUP's Annual Survey of Faculty Compensation.)

l°There exists the possibility that my estimates of the union-nonunion differentials might have been different from Freeman's if I had used identical controls. In order to explore this possibility, I re-estimated the equations in Tables 2, 3, and 4 without the variables Faculty Size, Faculty-Student Ratio, Large City, Small Town~RuraL and No Ranks, while adding a measure of student enrollment. (Even with these changes the~e were still minor differences between the controls used by myself and Freeman.) The re-estimated union- nonunion differentials tended to be 1 to 2 percentage points higher than those reported, but nevertheless fall well within the range of estimates presented by Freeman.

I IUsing mean salary as the dependent variable produced a similar pattern of results.

12A Chow test (at the .10 level) cannot reject the hypothesis that the coefficients of Union and Years Orga- nized are equal in the public and private sector samples. Also, a Chow test cannot reject the hypothesis that the coefficient of Union is equal across the two samples.

13A Chow test (at the. 10 level) cannot reject the hypothesis that the coefficients of Union and Years Orga- nized are equal in the two- and four-year school samples. Also, a Chow test cannot reject the hypothesis that the coefficient of Union is equal across the two samples.


14According to Douglas (1990, p. 106), 9.9 percent of faculty bargaining units were affiliated with the AAUP in 1989. The corresponding figure for the AFT was 29.9 percent; for the NEA, 48.5 percent; for independent or other national organizations, 10.5 percent; and for bargaining units with joint affiliation, 1.5 percent.

15An F-test (at the .10 level) rejects the hypothesis that the coefficients of the union affiliation variables are equal in the compensation and salary equations.

16The results of this section should be interpreted with caution. Faculty bargaining units frequently switch affiliations, and it may be the case that national unions actively seek to recruit strong bargaining units. Thus, it is possible that the differences in salary and compensation levels associated with national unions reflect only their relative abilities to woo faculty who have already unionized.

17This 15 percent figure refers to wages and salaries only. Estimates presented in Lewis (1986, pp. 96- 104) suggest that the average compensation differential may be 1 or 2 percentage points higher.

18Another way to approach the unobserved variable problem is to use Heckman's (1979) procedure. A reduced form probit was run in which the probability of a publicly controlled school being organized in 1987-1988 was specified to be a function of a set of variables that included a series of dummies denoting a state's legal environment for collective bargaining in higher education. An inverse Mills ratio was then calculated for each school and added to separate union and nonunion compensation equations. The inverse Mills ratio was not statistically significant in either equation, suggesting the absence of sample selection bias.

19Using mean salary as the dependent variable produced a similar pattern of results.

2°This perverse result may also be due to the violation the standard assumptions of the fixed-effects model (i.e., that the parameters of the model are constant with respect to time). Jakubson (1991, pp. 5-6) proposes a test of these assumptions. Using this test, I cannot reject the hypothesis that these assumptions are valid.

21Freeman also uses data for academic year 1965-1966 in his fixed effects analysis.

22NLRB v. Yeshiva University, 444 U.S. 672 (1980).

R E F E R E N C E S

Bognanno, Mario, David Estenson, and Edward Suntrup. "Union-Management Contracts in Higher Edu- cation." Industrial Relations 17 (May 1978): 189-203.

Bacharach, Samuel, Timothy P. Schmidle, and Scott C. Bauer. "Higher Education." In David B. Lipsky and Clifford B. Donn, eds., Collective Bargaining in American Industry. Lexington Mass: Lexington Books, 1987, pp. 225-64.

Barbezat, Debra A. "The Effect of Collective Bargaining on Salaries in Higher Education." Industrial and Labor Relations Review 42 (April 1989): 443-55.

Bimbaum, Robert. "Unionization and Faculty Compensation." Educational Record 55 (Winter 1974): 29-33.

"Unionization and Faculty Compensation: Part II." Educational Record 57 (Spring 1976): 116-18.

Brown, Charles. "Equalizing Differences in the Labor Market." Quarterly Journal of Economics 94 (February 1980): 113-134.

and James Medoff. "The Employer Size-Wage Effect." Journal of Political Economy 97 (October 1989): 1027-53.

Davis, Elizabeth E. A Fixed Effects Model of the Compensation Gains of Faculty Unions. Unpublished paper, available from the author at Abt Associates, 55 Wheeler St. Cambridge, MA 02138.


Douglas, Joel M. Directory of Faculty Contracts and Bargaining Agents in Institutions of Higher Educa- tion. New York: The National Center for the Study of Collective Bargaining in Higher Education and the Professions, 1990.

Freeman, Richard B. Should We Organize? Effects of Faculty Unionism on Academic Compensation. NBER Working Paper No. 301, 1978.

Gordon, Nancy, Thomas E. Morton, and Ina C. Braden. "Faculty Salaries: Is there Evidence Of Discrimi- nation By Sex, Race, and Discipline?" American Economic Review 64 (June 1974): 419-27.

Guthrie-Morse, Barbara, Larry Leslie, and Teh-Wei Hu. "Assessing the Impact of Faculty Unions: The Financial Impact of Collective Bargaining." Journal of Higher Education 52 (May/June 1981): 237-55.

Hansen, Lee W. "Changes in Faculty Salary." In Howard R. Bowen and Jack H. Schuster American Pro- fessors: A National Resource Imperiled. New York: Oxford University Press, 1986.

Heckman, James. "Sample Bias as a Specification Error." Econometrica 47 (January 1979): 153-62.

Hu, Teh-Wei, and Larry L. Leslie. "The Effects of Collective Bargaining on College Faculty Salaries and Compensation." Applied Economics 14 (June 1982): 269-77.

Jakubson, George. "Estimation and Testing of the Union Wage Effect Using Panel Data." Review of Eco- nomic Studies 58 (January 1991): 1-21.

Kesselring, Randall G. "The Economic Effects of Faculty Unions." Journal of Labor Research 12 (Winter 1991): 62-72.

Leslie, Larry L. and Teh-Wei Hu. "The Financial Implications of Collective Bargaining." Journal of Edu- cation Finance (Summer 1977): 32-52.

Lewis, H. Gregg. Union Relative Wage Effects: A Survey. Chicago: The University of Chicago Press, 1986.

Marshall, Joan. "The Effects of Collective Bargaining on Faculty Salaries in Higher Education." Journal of Higher Education 50 (May/June 1979): 310-22.

Mellow, William. "Employer Size and Wages." Review of Economics and Statistics 64 (August 1982): 495-501.

Morgan, David and Richard Kearny. "Collective Bargaining and Faculty Compensation: A Comparative Analysis." Sociology of Education 50 (January 1977): 28-39.

Wellington, Harry H. and Ralph K. Winter, Jr. "The Limits of Collective Bargaining in Public Employ- ment." Yale Law Journal 78 (June 1969): 1107-27.

the effect of unionization on faculty salaries and compensation: estimates from the 1980s

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