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These arguments notwithstanding, we believe that a logical starting point is to hypothesize that nonproŽt compensation is determined in competitive labor markets. Competition implies that the marginal worker will be indifferent between identical positions in nonproŽt and proŽt-seeking enterprises. In its absence, some jobs will be rationed and some employers will pay more than needed to fulŽll their demand for labor. However, competitive markets need not require identical levels of pay. As mentioned, wages may deviate if there are compensating differentials or if indi- viduals are willing to donate labor to nonproŽts. Therefore, we are particularly inter- ested in considering the joint hypothesis of competitive labor markets and the ab- sence of labor donations. The testable prediction is that the nonproŽt wage differential will be eliminated by including sufŽcient controls for worker and job characteristics.

III. Previous Research

Previous studies of nonproŽt compensation, summarized in Table 1, yield ambiguous results. Early examinations Johnston and Rudney 1987; Shackett and Trapani 1987; Preston 1989 suggest a large nonproŽt wage penalty but are hampered by the lack of information on the type of employer, requiring the imputa- tion of nonproŽt status. Researchers focusing on narrowly deŽned industries obtain equivocal Žndings. Weisbrod 1983 shows that public interest lawyers earn 20 percent less than those in the private sector and believes this is due to heterogeneity in preferences, rather than in worker quality. However, using the same data, Goddeeris 1988 claims the lower wages reect personal characteristics and that public interest attorneys earn no less than if employed by proŽt-seeking companies. Borjas, Frech III, and Gins- burg 1983 argue that the relatively high pay observed in nonproŽt nursing homes represents rent-sharing due to attenuated property rights. Conversely, Holtmann and Idson 1993 claim the wage premium occurs because nonproŽt nursing homes use higher quality labor and that registered nurses could actually earn more if they switched to for-proŽt facilities. Preston 1988 shows that federally regulated non- proŽt day care centers pay 5 to 10 percent more than for-proŽt facilities and interprets this as evidence of philanthropic wage-setting. However, she Žnds no differential for non-federally regulated centers. Mocan and Tekin forthcoming show that the size of the nonproŽt premium in this industry varies considerably with the type of ownership, characteristics of the staff, and hours worked. 12 Leete’s 2001 examination of data from the 1990 Census indicates that the overall nonproŽt differential is eliminated by including detailed controls for industries and occupations. Within three-digit industries, nonproŽt workers are as likely to obtain statistically signiŽcant wage premiums as penalties. These conclusions need to be interpreted with caution, however, because the controls for industries and occupa- 12. Roomkin and Weisbrod 1999 indicate that there is ambiguity even within industries. Focusing on six top managerial positions in hospitals, they Žnd lower nonproŽ t compensation in three chief executive ofŽcer, chief operating ofŽcer, and top patient care executive but higher pay in three others chief Žnancial ofŽcer, top human resources executive, and head of nursing services. h e Journal of H um an R es ourc es Table 1 Previous Research on NonproŽt Earnings Differentials Study Data Results Comments Borjas Frech III, and 1973–74 National Nursing Nursing home workers in religious- Many results are statistically insig- Ginsburg 1983 Home Survey afŽliated nonproŽts earn 4 percent niŽcant or sensitive to the choice less per hour than for-proŽt em- of speciŽcations. ployees; those in other nonproŽts receive an insigniŽcant 1.6 per- cent premium. Some evidence of higher wages for homes with more generous Medicaid reim- bursement programs. DuMond 1997 1995 Current Population NonproŽt workers earn 6 percent Not clear how government workers Survey Outgoing Rota- 11 percent less per hour than are treated. Small number of tran- tion Groups counterparts without with con- sitions between for-proŽt and non- trols for industry and occupation. proŽt employment in wage- Larger differential for males 19 change models. percent than females 0 to 5 per- cent. Gaps shrink to an insigniŽ- cant 0 to 4 percent in Žrst-differ- ence models. NonproŽt workers have higher pensionhealth insur- ance coverage and lower displace- ment rates. R uhm and B orkoski 997 Frank 1996 Cornell Employment Sur- NonproŽt differential in annual earn- Small and unrepresentative sample vey and other sources. ings was 259 percent for recent Cor- in main analysis; few controls. nell graduates, controlling for sex, GPA, and college curriculum. Other evidence of negative compensating differentials for working for socially responsible employers. Goddeeris 1988 Nationally representative Public Interest lawyers PIL earn Sector deŽnitions differ from Weis- surveys of private and pub- 37 percent less than those in private brod 1983. Selection identiŽed by lic interest lawyers in Žrms but this is entirely due to dif- community size, political activities 19734. ferences in characteristics. They orientation. would earn no less if they switched into the private sector. Holtmann and Idson Registered nurses in 1985 NonproŽts employ higher quality No distinction between government 1993 National Nursing Home registered nurses. OLS models re- and private nonproŽt nursing homes. Survey veal a 3 percent hourly wage pre- IdentiŽcation restrictions of selectiv- mium in nonproŽt homes and ity-corrected models are question- steeper experiencetenure proŽles for able. them. However, selectivity-corrected models indicate that nurses in non- proŽts actually earn less than they would if employed in for-proŽts. h e Journal of H um an R es ourc es Table 1 continued Study Data Results Comments Johnston and Rud- 1982 Census of service in- The average annual earnings of non- Hospitals, educational institutions, ney 1987 dustries proŽt workers are 21.5 percent and religious organizations ex- less than those employed in for- cluded. No controls for individual proŽt Žrms. characteristics. Leete 2001 1990 Census, 5 percent No overall nonproŽt wage differen- Estimated hourly wages may be sub- public use microdata tial after including detailed con- ject to measurement error. Ex- sample trols for industry and occupation. tremely detailed industry-occupa- Among speciŽed 3-digit industries tion interactions could absorb with statistically signiŽcant non- nonproŽt effects. proŽt differentials, positive and negative effects are equally likely. Mocan and Tekin 398 child care centers in NonproŽt childcare employees work- Extensive controls for human capital forthcoming Calif., Colorado, Conn., ing full-time part-time receive a and center characteristics. Discrete and N.C., data from 6 20 percent hourly wage pre- factor methods used to control for Spring 1993 mium and 8 10 percent higher to- unobserved heterogeneity. Stan- tal compensation. Considerable dard errors not reported. variation by type of nonproŽt and worker. R uhm and B orkoski 999 Preston 1988 Abt Associates, 1976–77 NonproŽt weekly wage premium of Center characteristics, labor quality, National Day Care Center 5 to 10 percent for childcare work- parental participation, and donations Supply Study ers in federally regulated daycare controlled for. Some differences centers; no difference for other cen- across center types could persist. ters. Results consistent with the for- mer being less competitive and able to pay rents to workers. Preston 1989 1990 Survey of Job Char- OLS results for SJC imply negative SJC sample is small n 300. Ex- acteristics SJC; May nonproŽt differential of 20 percent clusion restrictions are questionable 1979 Current Population for managersprofessionals, no effect for selectivity-corrected estimates. Survey CPS for clerical workers; larger negative NonproŽt status inferred not ob- effects for both groups in CPS. Se- served in CPS data. lectivity-corrected results sensitive to model speciŽcation. CPS wage change regressions indicate no differ- ential for clerical workers, statisti- cally insigniŽcant 10 percent pre- mium for managers and professionals. For-proŽt workers more often have pensions, health in- surance. Roomkin and Weis- Hay Management Consul- NonproŽt hospitals offer higher base Job complexity and hospital charac- brod 1999 tants, 1992 hospital com- salaries but lower bonus payments teristics controlled for; individual pensation survey to six top managerial positions. To- characteristics are not. Low response tal compensation is higher in three rate 19 percent. positions and lower in three. h e Journal of H um an R es ourc es Table 1 continued Study Data Results Comments Shackett and Tapani National longitudinal sur- Compared to private nonregulated in- NonproŽt status not observed; in- 1987 veys of young men and dustries, the nonproŽt wage differ- stead it is assumed to include all young women ential is 11, 0, 214, and 28 per- persons in hospital and educa- cent for white females, black tional services industries. females, white males, and black males. Weisbrod 1983 Same as Goddeeris 1988 PIL lawyers earn 20 percent less an- Small sample size 53 PIL lawyers; nually than if employed in private PIL lawyers may not be represen- sector. These attorneys are aware tative of other attorneys in non- of the negative earnings effects proŽts. Work hours and fringe and expect them to be permanent. beneŽts not controlled for. Differences in preferences consis- tent with type of employment. tions are so extensive as many as 20,000 industry-occupation interactions in some models that there is likely to be little variation in the type of employer within many of the narrowly deŽned industry-occupation cells. 13 Most similar to the present research is DuMond’s 1997 analysis of data from the 1994– 1995 Current Population Survey Outgoing Rotation Groups CPS-ORG. His cross-sectional regressions indicate a nonproŽt wage penalty of between 6 and 11 percent. Conversely, Žxed-effect estimates, exploiting data on individuals switching between for-proŽt and nonproŽt employment, imply small 0 to 4 percent and statis- tically insigniŽcant earnings gaps. Several factors reduce our conŽdence in these Žndings. First, it is not clear how movements into or out of public sector are treated. Second, few respondents switch types of employment over the two-year period, de- creasing the precision of the estimates. Third, endogenous mobility between sectors is not considered. Fourth, DuMond controls only for broad one-digit industries or occupations, which might inadequately account for differences in the job characteris- tics of nonproŽt and for-proŽt employment. Each of these issues receives attention below.

IV. Data