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Wages and performance-based pay: Evidence from the NLSY
economics letters ELSEVIER
Economics Letters 51 (1996) 241-246
Wages and performance-based pay: Evidence from the NLSY B r a d l e y T. E w i n g Department of Finance and Economics, Landrum Box 8151, Georgia Southern University, Statesboro, GA 30460-8151, USA Received 8 May 1995; accepted 30 October 1995
Abstract
The link between performance and pay should be strongest where performance is more accurately observed, Brown (Industrial and Labor Relations Review, 1990, 43, 1655-1825; RAND Journal of Economics, 1992, 23, 366-375). More productive workers self-select into jobs with performance-based pay. Consequently, workers whose pay is based on performance should earn more than those whose pay is not based on such measures. This paper adds to the literature on the subject by providing new empirical evidence for Brown's model using the National Longitudinal Surveys of Youth (NLSY) data.
Keywords: Pay; Wages; Performance; NLSY JEL classification: J33
I. Introduction
This paper tests the claim that workers whose pay is based on performance earn more than those whose pay is not based on such measures. Brown's (1990, 1992) simple, but elegant, model of worker compensation predicts that the link between pay and performance is stronger where performance is more accurately observed. Brown focuses on three methods of pay: piece-rate, merit, and standard rate. In his model, a firm using piece-rates directly ties the worker's performance to her compensation. The merit rate firm indirectly links pay to performance through some type of evaluation system, such as a supervisor's ratings. Standard rate firms do not link pay to performance. Brown shows that the average pay of piece-rate workers should be greater than that of merit rate workers, which should be greater than that of standard rate workers. His empirical tests of manufacturing workers' earnings show that while piece-rate workers earn the most, standard rate workers earn more than merit rate workers. The purpose of this paper is to examine the p a y - p e r f o r m a n c e link. We concentrate on types 0165-1765/96/$12.00 (~) 1996 Elsevier Science S.A. All rights reserved SSDI 0165-1765(95)00775-X
242
B.T. Ewing / Economics Letters 51 (1996) 241-246
of performance-based pay used by many firms and for many different jobs. The use of a different type of data set (self-reported vs. establishment level) will provide insight into the robustness of the hypothesis. The data provide a rich set of control variables, enabling us to focus on the effect of the method of pay fairly directly. We find evidence consistent with Brown's theory. In particular, workers whose pay is based on performance earn a substantial wage premium.
2. The data
The data come from the National Longitudinal Surveys of Youth (NLSY) which has interviewed respondents annually from 1979 to the present. The initial wave contained 12,686 individuals between the ages of 14 and 21. Our sample consists of persons who worked for pay in the year prior to the 1990 wave. See Table 1 for variable definitions and descriptive statistics. The 'method of pay' variables are based on response to the questions of whether or not earnings are based on performance, bonuses, or commissions. By construction, 'performance pay' includes both bonus and commissions, as well as other types of performance-based pay. Profit-sharing and employee stock-purchase plans are not included in the definition of performance-based pay. Use of self-reported data differs from that of Brown, who uses the BLS Industry Wage Survey, a firm/establishment data set. Consequently, a finding of the p e r f o r m a n c e - p a y link is all the more supportive of Brown's theory. It is possible with the NLSY to construct a measure of work experience that represents actual weeks worked, less tenure at the current firm. There are several reasons why a measure of actual experience is preferred to using potential work experience (usually defined as age-education-6). Potential experience may understate the returns to experience because it treats time not working in the same way as time working. This is particularly troublesome when estimating the wages of females who are more likely to have intermittent labor force participation than are males. 1 The use of both actual experience and tenure at the current firm should capture the total work experience of the respondent. There are several reasons for including the A r m e d Forces Qualifications Test ( A F Q T ) in the wage model. First, it may proxy for unobserved ability (Blackburn and Neumark, 1992). Secondly, Maxwell (1994) has successfully argued that A F Q T proxies for quality of schooling received. We include A F Q T in our regressions in addition to years of education. In this respect we incorporate elements of both school quality and quantity. The monitoring of workers by firms is a recurrent theme in Brown's papers. Therefore, we include a vector of firm-specific variables that includes establishment size, a control for whether the employer has multiple locations, and for collective bargaining coverage. These variables, along with industry and occupation controls, presumably capture much of the heterogeneity in monitoring technology. An additional variable that indicates whether the
1The timing of work interruptions may affect the degree of human capital depreciation (Stratton, 1995). We do not consider this a problem in our study as the evidence indicates it may be negligible.
B.T. Ewing / Economics Letters 51 (1996)241-246
243
Table 1 Selected variable definitions and descriptive statistics Variable
Mean (std. error)
In Wage
= natural log of wage
2.1727 (0.4065)
Performance pay
= 1 if pay is based on performance
0.2088 (0.4065)
Bonus pay
= 1 if pay is based on bonus
0.1309 (0.3374)
Commissions
= 1 if pay is based on commissions
0.0596 (0.2369)
Experience
= actual weeks worked less tenure
285.67 (151.12)
Tenure
=weeks at current employer
160.69 (148.35)
Full time
= 1 if works >~35 hours per week
0.9553 (0.2068)
Collbarg
= 1 if covered by collective bargaining
0.1876 (0.3905)
agreement Multiple locations
= 1 if employer has multiple locations
Size
= establishment size
0.6779 (0.4674)
551.96 (2409.2)
Supervisor
= 1 if worker supervises others
0.3746 (0.4841)
n = 3,353
worker supervises others is also included. Other variables include controls for marital status, number of children in the household, race, and gender, as well as residence in a region of the country, urban area and SMSA. Our sample contains 3,353 persons, 53% of whom are male. Nearly 21% of the individuals have pay which is based on performance. Bonuses are linked to about 13% of the respondents' compensation, while 6% have earnings based on commissions. Workers have an average of 5.5 years of actual experience exclusive of time spent at their current employer. Average tenure is slightly more than 3 years. Almost all are full-time workers, due in part to the fact that to be included in our sample they had to answer many questions about their
244
B.T. Ewing / Economics Letters 51 (1996) 241-246
employer, actual weeks worked in successive years, etc. It is reasonable to assume that part-time workers might know less about their employer than do full-time workers. Consequently, our estimates may suffer from a sample selection bias that arises from the decision to work full time. However, experiments with sample selection models did not suggest this was a problem. In all cases the coefficient on the selection term was insignificant. Therefore, we decided to estimate the model without a sample selection term, and to include a d u m m y variable for full-time work. Nearly 19% of our sample is covered by collective bargaining agreements; 68% work at firms with more than one location, and the typical establishment has about 552 employees. A b o u t 37% of the workers direct the work of others. Finally, slightly more than half of the workers are married and the average number of children in the household is one.
3. Results Results of the log wage regressions are presented in Table 2. Column I indicates that performance-based pay workers earn a 5.5% wage premium. If the performance-based pay takes the form of bonuses, the premium is around 4% (column II), while those whose pay has a commissions aspect to it earn about 9% more than non-performance-based workers (column III). Estimated coefficients on the independent variables in the three model specifications are very similar and consistent with the literature on wage determination. In the interest of brevity we present a few of the more interesting findings. First, tenure behaves as expected. Additional weeks worked produce higher pay, but at a decreasing rate. The coefficient on actual experience is positive but insignificant. This may be because younger workers' experience, net of that at the current firm, is not considered that valuable. Perhaps young workers accept low-paying service sector jobs at the beginning of their careers that do not provide much training. However, the coefficient on the squared term is positive and significant, suggesting that there may be some threshold level of experience required before it is rewarded in the labor market. Workers employed by larger firms, and by firms with multiple locations, have bigger wages. This finding suggests that where monitoring is more difficult, workers are paid more. Finally, being in a position of supervising others brings with it about a 7% increase in pay.
4. Concluding remarks This paper examined the effect that performance-based pay has on earnings. The results support Brown's contention that the pay-performance link is stronger where performance is more accurately observed. Having controlled for many personal, institutional, firm-specific, and demographic factors that affect the wage, we find that workers who are in performancebased pay jobs earn a substantial wage premium. This result is robust across several measures of performance-based pay.
B.T. Ewing / Economics Letters 51 (1996) 241-246
245
Table 2 Log wage model I
Performance pay
II
III
0.0552 (3.407)
Bonus pay
0.0395 (2.059)
Commissions
0.0911 (3.192)
Education
0.0363 (9.097)
0.0363 (9.079)
Experience
0.0001 (0.767)
0.0001
0.0002
(0.725)
(0.889)
Experience 2
0.707 * 10 -6 (2.313)
0.720 * 10 -6 (2.358)
0.661 * 10 -6 (2.163)
Tenure
0.0016 (10.902)
0.0017 (10.964)
0.0016 (10.908)
Tenure 2
-0.138.10 5 (4.622)
-0.139 * 10 -5 (4.718)
0.0362 (9.066)
- 0 . 1 3 6 . 1 0 -5 (4.609)
Collbarg
0.1379 (7.815)
0.1349 (7.654)
0.1360 (7.723)
Multiple locations
0.0557 (3.914)
0.0546 (3.835)
0.0564 (3.959)
Size
0.878 * 10 -5 (3.285)
0.875 * 10 -5 (3.268)
0.914 * 10-5 (3.418)
AFQT
0.0026 (8.140)
0.0026 (8.170)
0.0026 (8.151)
Supervisor
0.0709 (5.003)
0.0701 (2.059)
0.0728 (5.144)
Constant
0.6637 (9.926)
0.6780 (10.134)
0.6699 (10.024)
0.4781
0.4770
0.4779
adjusted R2
Note: Absolute t-ratios in parentheses. Additional variables included controls for industry, occupation, full time, martial status, number of children in the household, race, gender, region of country, SMSA, and urban residence. Full results available from the author upon request.
246
B.T. Ewing / Economics Letters 51 (1996) 241-246
References Blackburn, M. and D. Neumark, 1992, Unobserved ability, efficiency wages, and interindustry wage differentials, Quarterly Journal of Economics 107, 1421-1436. Brown, C., 1990, Firms' choice of method of pay, Industrial and Labor Relations Review 43, 165S-182S. Brown, C., 1992, Wage levels and method of pay, RAND Journal of Economics 23, 366-375. Maxwell, N., 1994, The effect on black-white wage differences of differences in the quantity and quality of education, Industrial and Labor Relations Review 47, 249-263. Stratton, L., 1995, The effects that interruptions in work experience have on wages, Southern Economic Journal 61, 955-970.
Economics Letters 51 (1996) 241-246
Wages and performance-based pay: Evidence from the NLSY B r a d l e y T. E w i n g Department of Finance and Economics, Landrum Box 8151, Georgia Southern University, Statesboro, GA 30460-8151, USA Received 8 May 1995; accepted 30 October 1995
Abstract
The link between performance and pay should be strongest where performance is more accurately observed, Brown (Industrial and Labor Relations Review, 1990, 43, 1655-1825; RAND Journal of Economics, 1992, 23, 366-375). More productive workers self-select into jobs with performance-based pay. Consequently, workers whose pay is based on performance should earn more than those whose pay is not based on such measures. This paper adds to the literature on the subject by providing new empirical evidence for Brown's model using the National Longitudinal Surveys of Youth (NLSY) data.
Keywords: Pay; Wages; Performance; NLSY JEL classification: J33
I. Introduction
This paper tests the claim that workers whose pay is based on performance earn more than those whose pay is not based on such measures. Brown's (1990, 1992) simple, but elegant, model of worker compensation predicts that the link between pay and performance is stronger where performance is more accurately observed. Brown focuses on three methods of pay: piece-rate, merit, and standard rate. In his model, a firm using piece-rates directly ties the worker's performance to her compensation. The merit rate firm indirectly links pay to performance through some type of evaluation system, such as a supervisor's ratings. Standard rate firms do not link pay to performance. Brown shows that the average pay of piece-rate workers should be greater than that of merit rate workers, which should be greater than that of standard rate workers. His empirical tests of manufacturing workers' earnings show that while piece-rate workers earn the most, standard rate workers earn more than merit rate workers. The purpose of this paper is to examine the p a y - p e r f o r m a n c e link. We concentrate on types 0165-1765/96/$12.00 (~) 1996 Elsevier Science S.A. All rights reserved SSDI 0165-1765(95)00775-X
242
B.T. Ewing / Economics Letters 51 (1996) 241-246
of performance-based pay used by many firms and for many different jobs. The use of a different type of data set (self-reported vs. establishment level) will provide insight into the robustness of the hypothesis. The data provide a rich set of control variables, enabling us to focus on the effect of the method of pay fairly directly. We find evidence consistent with Brown's theory. In particular, workers whose pay is based on performance earn a substantial wage premium.
2. The data
The data come from the National Longitudinal Surveys of Youth (NLSY) which has interviewed respondents annually from 1979 to the present. The initial wave contained 12,686 individuals between the ages of 14 and 21. Our sample consists of persons who worked for pay in the year prior to the 1990 wave. See Table 1 for variable definitions and descriptive statistics. The 'method of pay' variables are based on response to the questions of whether or not earnings are based on performance, bonuses, or commissions. By construction, 'performance pay' includes both bonus and commissions, as well as other types of performance-based pay. Profit-sharing and employee stock-purchase plans are not included in the definition of performance-based pay. Use of self-reported data differs from that of Brown, who uses the BLS Industry Wage Survey, a firm/establishment data set. Consequently, a finding of the p e r f o r m a n c e - p a y link is all the more supportive of Brown's theory. It is possible with the NLSY to construct a measure of work experience that represents actual weeks worked, less tenure at the current firm. There are several reasons why a measure of actual experience is preferred to using potential work experience (usually defined as age-education-6). Potential experience may understate the returns to experience because it treats time not working in the same way as time working. This is particularly troublesome when estimating the wages of females who are more likely to have intermittent labor force participation than are males. 1 The use of both actual experience and tenure at the current firm should capture the total work experience of the respondent. There are several reasons for including the A r m e d Forces Qualifications Test ( A F Q T ) in the wage model. First, it may proxy for unobserved ability (Blackburn and Neumark, 1992). Secondly, Maxwell (1994) has successfully argued that A F Q T proxies for quality of schooling received. We include A F Q T in our regressions in addition to years of education. In this respect we incorporate elements of both school quality and quantity. The monitoring of workers by firms is a recurrent theme in Brown's papers. Therefore, we include a vector of firm-specific variables that includes establishment size, a control for whether the employer has multiple locations, and for collective bargaining coverage. These variables, along with industry and occupation controls, presumably capture much of the heterogeneity in monitoring technology. An additional variable that indicates whether the
1The timing of work interruptions may affect the degree of human capital depreciation (Stratton, 1995). We do not consider this a problem in our study as the evidence indicates it may be negligible.
B.T. Ewing / Economics Letters 51 (1996)241-246
243
Table 1 Selected variable definitions and descriptive statistics Variable
Mean (std. error)
In Wage
= natural log of wage
2.1727 (0.4065)
Performance pay
= 1 if pay is based on performance
0.2088 (0.4065)
Bonus pay
= 1 if pay is based on bonus
0.1309 (0.3374)
Commissions
= 1 if pay is based on commissions
0.0596 (0.2369)
Experience
= actual weeks worked less tenure
285.67 (151.12)
Tenure
=weeks at current employer
160.69 (148.35)
Full time
= 1 if works >~35 hours per week
0.9553 (0.2068)
Collbarg
= 1 if covered by collective bargaining
0.1876 (0.3905)
agreement Multiple locations
= 1 if employer has multiple locations
Size
= establishment size
0.6779 (0.4674)
551.96 (2409.2)
Supervisor
= 1 if worker supervises others
0.3746 (0.4841)
n = 3,353
worker supervises others is also included. Other variables include controls for marital status, number of children in the household, race, and gender, as well as residence in a region of the country, urban area and SMSA. Our sample contains 3,353 persons, 53% of whom are male. Nearly 21% of the individuals have pay which is based on performance. Bonuses are linked to about 13% of the respondents' compensation, while 6% have earnings based on commissions. Workers have an average of 5.5 years of actual experience exclusive of time spent at their current employer. Average tenure is slightly more than 3 years. Almost all are full-time workers, due in part to the fact that to be included in our sample they had to answer many questions about their
244
B.T. Ewing / Economics Letters 51 (1996) 241-246
employer, actual weeks worked in successive years, etc. It is reasonable to assume that part-time workers might know less about their employer than do full-time workers. Consequently, our estimates may suffer from a sample selection bias that arises from the decision to work full time. However, experiments with sample selection models did not suggest this was a problem. In all cases the coefficient on the selection term was insignificant. Therefore, we decided to estimate the model without a sample selection term, and to include a d u m m y variable for full-time work. Nearly 19% of our sample is covered by collective bargaining agreements; 68% work at firms with more than one location, and the typical establishment has about 552 employees. A b o u t 37% of the workers direct the work of others. Finally, slightly more than half of the workers are married and the average number of children in the household is one.
3. Results Results of the log wage regressions are presented in Table 2. Column I indicates that performance-based pay workers earn a 5.5% wage premium. If the performance-based pay takes the form of bonuses, the premium is around 4% (column II), while those whose pay has a commissions aspect to it earn about 9% more than non-performance-based workers (column III). Estimated coefficients on the independent variables in the three model specifications are very similar and consistent with the literature on wage determination. In the interest of brevity we present a few of the more interesting findings. First, tenure behaves as expected. Additional weeks worked produce higher pay, but at a decreasing rate. The coefficient on actual experience is positive but insignificant. This may be because younger workers' experience, net of that at the current firm, is not considered that valuable. Perhaps young workers accept low-paying service sector jobs at the beginning of their careers that do not provide much training. However, the coefficient on the squared term is positive and significant, suggesting that there may be some threshold level of experience required before it is rewarded in the labor market. Workers employed by larger firms, and by firms with multiple locations, have bigger wages. This finding suggests that where monitoring is more difficult, workers are paid more. Finally, being in a position of supervising others brings with it about a 7% increase in pay.
4. Concluding remarks This paper examined the effect that performance-based pay has on earnings. The results support Brown's contention that the pay-performance link is stronger where performance is more accurately observed. Having controlled for many personal, institutional, firm-specific, and demographic factors that affect the wage, we find that workers who are in performancebased pay jobs earn a substantial wage premium. This result is robust across several measures of performance-based pay.
B.T. Ewing / Economics Letters 51 (1996) 241-246
245
Table 2 Log wage model I
Performance pay
II
III
0.0552 (3.407)
Bonus pay
0.0395 (2.059)
Commissions
0.0911 (3.192)
Education
0.0363 (9.097)
0.0363 (9.079)
Experience
0.0001 (0.767)
0.0001
0.0002
(0.725)
(0.889)
Experience 2
0.707 * 10 -6 (2.313)
0.720 * 10 -6 (2.358)
0.661 * 10 -6 (2.163)
Tenure
0.0016 (10.902)
0.0017 (10.964)
0.0016 (10.908)
Tenure 2
-0.138.10 5 (4.622)
-0.139 * 10 -5 (4.718)
0.0362 (9.066)
- 0 . 1 3 6 . 1 0 -5 (4.609)
Collbarg
0.1379 (7.815)
0.1349 (7.654)
0.1360 (7.723)
Multiple locations
0.0557 (3.914)
0.0546 (3.835)
0.0564 (3.959)
Size
0.878 * 10 -5 (3.285)
0.875 * 10 -5 (3.268)
0.914 * 10-5 (3.418)
AFQT
0.0026 (8.140)
0.0026 (8.170)
0.0026 (8.151)
Supervisor
0.0709 (5.003)
0.0701 (2.059)
0.0728 (5.144)
Constant
0.6637 (9.926)
0.6780 (10.134)
0.6699 (10.024)
0.4781
0.4770
0.4779
adjusted R2
Note: Absolute t-ratios in parentheses. Additional variables included controls for industry, occupation, full time, martial status, number of children in the household, race, gender, region of country, SMSA, and urban residence. Full results available from the author upon request.
246
B.T. Ewing / Economics Letters 51 (1996) 241-246
References Blackburn, M. and D. Neumark, 1992, Unobserved ability, efficiency wages, and interindustry wage differentials, Quarterly Journal of Economics 107, 1421-1436. Brown, C., 1990, Firms' choice of method of pay, Industrial and Labor Relations Review 43, 165S-182S. Brown, C., 1992, Wage levels and method of pay, RAND Journal of Economics 23, 366-375. Maxwell, N., 1994, The effect on black-white wage differences of differences in the quantity and quality of education, Industrial and Labor Relations Review 47, 249-263. Stratton, L., 1995, The effects that interruptions in work experience have on wages, Southern Economic Journal 61, 955-970.