BLS WORKING PAPERS 

 

U.S. Department of Labor 

U.S. Bureau of Labor Statistics 

Office of Compensation and Working Conditions 

Some New Evidence on Overtime Use, Total Job Compensation, and 

Wage Rates 

Anthony J. Barkume, U.S. Bureau of Labor Statistics 

Working Paper 402

January 2007

All views expressed in this paper are those of the author and do not necessarily reflect the views or policies of the 

U.S. Bureau of Labor Statistics. 
 

Some New Evidence on Overtime Use, Total Job Compensation, and Wage Rates 

Anthony J. Barkume*

Abstract: This paper is a replication of research reported by Steven Trejo in the 1991 

American Economic Review. Trejo used labor force data from the seventies to assess the 

relevance of two contrasting views of the impact of overtime pay regulation. This paper 

reports research using a recent representative sample of U.S. private industry jobs that 

includes employer-reported measures of usual annual hours of overtime work and 

comprehensive measures of employer costs for job compensation. Comparisons are made 

between a set of jobs likely to be subject to U.S. overtime pay regulationójobs paid 

hourly on 40 hour a week schedulesówith another set of jobs that can offer overtime but 

are not likely to be subject to Federal overtime requirementsójobs on reduced hour 

schedules. The main findings of the research are: (1) higher wage rates are associated 

with a lower incidence of overtime work among the set of jobs with 40 hour work 

schedules, but not among the set of jobs with reduced hour schedules (2) in jobs using 

overtime work, more usual overtime work is associated with lower wage rates among the 

jobs with 40 hour work schedules, but not among the jobs on reduced hour schedules (3) 

higher ìquasi-fixedî job compensation, such as employer health insurance costs, is 

associated with a higher incidence of overtime use. The paper also discusses some of the 

difficulties of interpreting these statistical results in the context of the labor market 

models considered by Trejo. 

JEL Codes: J33 (Compensation Packages; Payment Methods); J31 (Wage Level and 

Structure; Wage Differentials) 

Keywords: Overtime work hours; Hedonic Wage Curve 

 

*

 Research Economist, U.S. Bureau of Labor Statistics; e-mail: barkume.anthony@bls.gov. John Bishow 

provided substantial and excellent research support on this project. Steven Trejo provided useful comments 

on an earlier version of the paper. The data used for this study have confidentiality restrictions, but for 

those interested in replication of the results, please see footnote 11. The views expressed in this paper are 

those of the author and do not reflect the policies of the U.S. Bureau of Labor Statistics. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 1

Introduction

Most economic analysis of hours worked focuses on the demand and supply for hours 

worked--how optimal hours of work for firms and workers depend on an hourly price of labor. 

The usual textbook analysis of overtime pay regulation takes a labor demand approach; see, for 

example, Ehrenberg and Smith (2000, pp. 150-153). In the labor demand model, the requirement 

of an overtime premium increases the price of labor in the overtime region, thereby reducing the 

firmís demand for hours and encouraging substitution of workers for overtime hours in the 

composition of the firmís labor input. In contrast, Lewis (1969) and Barzel (1973) developed a 

different view of the employment relationship in which a statutory overtime premium has no 

necessary impacts on the employment bargain between workers and firms. In particular, Trejo 

(1991) observed that in the Lewis-Barzel analysis, the preferred ìjob packageî of total earnings 

and hours worked will be unaffected by an overtime pay requirement so long as workers and 

firms can agree on compensating adjustments in the straight time wage rate or other 

compensation. 

How the adjustments in the structure of compensation in the Lewis-Barzel analysis works 

in the context of U.S. overtime pay regulation, which requires payment of a premium of one and a 

half of the usual straight-time wage rate beyond forty hours a week in most hourly jobs, can be 

illustrated with a simple example. If workers had bargained with a firm to work 50 hours a week 

at a wage of $6 per hour, the ìpackageî of weekly hours and earnings ($300 per week for 50 

hours of work) would be unaffected by imposition of an overtime premium if, for example, the 

payment of weekly earnings was restructured to account for the overtime premium, such lowering 

the straight-time wage rate to $5 per hour and adding a ìweekly bonusî of $25. Besides 

transactions costs, an important barrier to making such adjustments in low wage jobs is the 

minimum wage; all jobs subject to overtime pay requirements are also subject to the minimum 

wage under the U.S. Fair Labor Standards Act (FLSA). Costa (2000) found that at the time of the 

enactment of the FLSA in 1938, which introduced both the Federal minimum wage and the 

overtime pay requirement, overtime hours worked in Southern retail and wholesale trade 

industries declined; Costa attributed this decline to the low prevailing wages in the South prior to 

the enactment of the FLSA. 

Trejo (1991) analyzed several cross sections of Current Population Survey (CPS) data on 

weekly earnings and usual hours worked to assess whether the traditional labor demand analysis, 

which he termed the Fixed-Wage Model, or the Lewis-Barzel analysis, which he termed the 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 2

Fixed-Job Model (since this model does not consider worker-overtime hour substitution 

explicitly1

), better predicts the pattern of overtime incidence and the structure of labor earnings 

observed among workers subject to U.S overtime pay regulation. Section II of the paper discusses 

more fully his methodology and findings, but here we review a few key results in Trejoís findings 

to better motivate the research reported in this paper. Trejo found the incidence of workers 

having exactly 40 hour weekly work schedules was higher in jobs with FLSA overtime pay 

regulation coverage, and Trejo interpreted this spike in the weekly hour distribution as consistent 

with the Fixed-Wage Model predicting lower incidence with overtime pay regulation. Workers in 

jobs paying the minimum wage were more likely to work exactly 40 hours a week (i.e., working 

less overtime), suggesting that the minimum wage acts as a barrier to the restructuring of 

compensation with overtime pay regulation, as predicted by the Fixed-Job Model. Other findings 

were consistent with the predictions of the Fixed-Job Model, but the degree of consistency of 

these findings depended on the measurement of earnings used. The cross section relationship 

between usual weekly earnings and overtime hours was exactly as predicted by the Fixed-Job 

Model, but the cross section relationship between the straight-time wage rate and overtime hours 

indicated only partial adjustment in the structure of labor earnings that would be required to 

equalize labor earnings between jobs requiring an overtime premium and jobs that did not. 

The purpose of this paper is to report on a replication of a large part of the research 

reported in Trejo (1991), using establishment data that provide more consistent and 

comprehensive data on worker compensation and a measure of overtime hours that is more 

appropriate for testing the predictions of the Fixed-Job Model on the impact of U.S. overtime pay 

regulation. The Fixed-Job Model is best seen as a proposition about the nature of the employment 

bargain with expected (ex ante) overtime hours. The firmís requirements for overtime can 

fluctuate from week to week due to expected variation of market demand or the timing of 

production so that one would expect in most employment situations that transactions costs would 

limit the adjustment of straight-time wage rates with weekly fluctuations in overtime hours. For 

the research completed for this paper, I use micro data from the U.S. Bureau of Labor Statistics 

National Compensation Survey (NCS), which is a representative sample of U.S. nonagricultural 

private industry jobs. Of key importance for the objectives of the current research is that the NCS 

collects data on usual annual overtime hours along with collection of comprehensive annual labor 

costs for employee compensation. We use a recent cross section of NCS data (March 2004) to 

analyze the relationship between annual overtime hours with the level and structure of usual 

 

1

 However, Cutler and Madrian (1998) use a variant of this model to analyze worker-hour substitution. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 3

annual labor compensation. In addition to examining the relationship between the amount of 

overtime used and adjustments in the straight-time wage rate that can be attributable to overtime 

pay regulation, I also examine the relationship between the amount of overtime hours and labor 

costs not directly related to the amount of overtime hours worked (such as the employer costs for 

health insurance contributions) because such ìquasi-fixedî costs of labor could be an alternative 

channel for the adjustments in the structure of job compensation induced by overtime pay 

regulation. Section IV of the paper discusses the NCS data set and the subsets of these data used 

in the analysis. 

Another difference between Trejoís original 1991 study and the current paper is an 

attempt in this paper to infer more about the effects of overtime pay regulation from variation in 

overtime hours and earnings within the group of jobs subject to overtime pay regulation, as 

compared to such variation within a group of jobs not subject to overtime pay regulation. The 

Lewis-Barzel analysis underlying the Fixed-Job Model uses a hedonic locus of total earnings and 

hours worked as an analytical point of departure. Section III of the paper reviews this connection 

in more detail as a foundation for the statistical analysis. Drawing on the analysis of Kinoshita 

(1987), we use the distinction between the statutory overtime premiumófor example, time and a 

half in U.S. overtime pay regulationófrom a ìJob Market Overtime Premiumî, a term I use to 

refer to the labor cost differential in an alternative job requiring an additional hour of overtime, to 

analyze the relationship between the level and structure of labor costs under an overtime pay 

constraint in the Fixed-Job Model. 

Sections V and VI present the methodology and results of the statistical analysis of the 

NCS data. Consistent with Trejoís findings, I conclude from an analysis of incidence that 

overtime pay regulation does introduce a kink in a jobís labor cost structure with overtime, 

consistent with the Fixed-Wage Model. Also, I find that full-time jobs paying at or close to the 

minimum wage and subject to overtime pay regulation do not have a lower incidence of overtime 

work. This result is contrary to Trejoís findings, which were based on data from the seventies 

when the real value of the minimum wage was much higher than in the current decade. 

Estimating a cross section relationship between total ìhours-relatedî labor costs and overtime 

hours worked, I find that the predicted log differential with overtime hours in these hours-worked 

driven labor costs is about 85 per cent of the labor costs predicted by the Fixed-Wage Model. I 

also show that this mitigation of the labor cost impact of U.S. overtime pay regulation occurs 

because of reductions in the straight-time wage rate with increased overtime, as implied by the 

Fixed-Job Model. Section VII concludes with some qualifications about making such inferences 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 4

about the empirical relevance of the Fixed-Job Model, given the study methodology, and suggests 

some areas of additional research. 

II. A Closer Look at the Trejo (1991) Research on Overtime Pay

Trejoís empirical strategy was to compare hours and earnings in jobs with jobs presumed 

to have the FLSA overtime requirements with those jobs presumed to be exempt from FLSA 

overtime requirements. In general there are three types of nonagricultural2

 jobs exempt from the 

FLSA overtime pay requirements: (1) salaried jobs that have job duties that meet certain tests for 

exemption, such as managers and executives3

 (2) some hourly paid occupations in transportation 

industries involved in interstate commerce that have overtime pay regulated by other Federal law, 

and (3) ìmiscellaneous exemptions that apply to certain workers in retail trade, services, and state 

and local governmentî (Trejo (1991), p. 724). Trejo used the worker population in groups (2) and 

(3)óworkers paid hourlyóas his comparison group. 

Two margins of firm decision-making that Trejo examined to obtain evidence on the 

effect of overtime pay regulation were the decision to offer overtime and, if overtime is offered, 

the earnings-hours combination offered in the job.4

 Since the Fixed-Wage Model implies a kink 

in the earnings-hours combination at 40 hours for firms subject to FLSA overtime pay regulation, 

incidence should be lower than in firms exempt from FLSA coverage. The only prediction for 

incidence of the Fixed-Job Model would be if minimum wage requirements (also dictated by the 

FLSA) prevented adjustments in the structure of overtime to realize a desired earnings-hours 

combination. For example, if a job exempt from FLSA overtime pay requirements paid $5.15 per 

hour and had a 44 hour work week, a comparable job covered by FLSA overtime pay 

requirements could not provide the same hours-earnings package because the requirement to pay 

time and a half would also require a base wage of $4.92 per hour, below the minimum wage of 

$5.15. 

Examining the incidence of overtime, Trejo did find a higher proportion of FLSA exempt 

jobs had work weeks greater than forty hours. (For this analysis, he used actual rather than usual 

work hour responses.) For example, in the 1974 data, 68 % of the workers working 40 or more 

hours in jobs covered by FLSA overtime requirements worked exactly 40 hours while only 55% 

 

2

 Agricultural jobs are also exempt, possibly because of the role hours worked plays in agricultural 

technology processes required to supply perishable products, such as in harvesting. 

3

 Since August 2004, job earnings thresholds have been added to these ìjob testsî criteria to define FLSA 

exempt status, but these changes occurred after the data used in this paper were collected. 

4

 Trejo (1991) also examines estimates of firm compliance with overtime pay regulation; this evidence is 

not considered in the present study. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 5

of workers in exempt jobs did so. As predicted by the Fixed-Wage Model the estimated 

probability of overtime work was lower in jobs with FLSA overtime coverage in two of the three 

cross sections of data studied. However, among workers with FLSA overtime coverage and 

working at least 40 hours a week, those earning minimum wage less likely to work overtimeó

more likely to work exactly 40 hours, as predicted by the Fixed-Job Model. 

To examine the effect on the structure of earnings for workers that did work overtime, 

Trejo noted that since the Fixed-Job Model assumes equalization of total earnings between any 

two equivalent jobs, the relative straight-time wage between a job exempt from the FLSA and 

another equivalent job subject to FLSA overtime pay regulation depends on the amount of 

overtime worked. Let W be the straight time wage rate in the FLSA-exempt job and Wí be the 

wage with FLSA coverage. With H denoting the common hours of work, with H > 40, then 

equalization of labor earnings between the two jobs implies W H = Wí [40 + 1.5 (H ñ 40)] = Wí 

[H + 0.5 (H ñ 40)]; thus (W / Wí) =1 + 0.5 [(H ñ 40)/H]. (For example, the differential in straight 

time pay of the FLSA exempt job would be 1.0235 with a 42 hour work week and 1.045 with a 44 

hour work week.) In a log straight-time wage regression on cross section data that controlled for 

other wage-determining characteristics the Fixed-Job Model would then predict5

 a coefficient of -

0.5 on a variable that interacts FLSA coverage with [(H ñ 40)/H]. Similarly, the Fixed-Job Model 

predicts a coefficient of zero on that variable in a log weekly earnings equation. 

For the log weekly earnings equation, Trejo measured as the dependent variable the CPS 

respondentís estimate of usual weekly earnings, although he noted6

 that respondents might not 

count usual overtime pay in reporting weekly earnings. Trejo found the effect of the ratio of 

overtime hours to total hours worked among jobs covered by FLSA on weekly earnings not to be 

statistically significant from zero, completely consistent with the predictions of the Fixed-Job 

Model. However using the CPS wage rate dataódata reported separately by CPS respondents 

from usual weekly earningsóthe magnitude of the overtime premium regulatory effect was much 

lower than as predicted by the Fixed-Job modelóabout a 0.2 differential in most estimations 

instead of the predicted 0.5 differential. (For example, with a 44 hour work week the straight-time 

wage rate equation would predict a relative straight time pay differential in exempt jobs of 1.018 

instead of 1.045.) Thus, these two (independent) data sources on job compensation did not give 

completely consistent results for the effects of overtime pay regulation on earnings. 

 

5

 The following is a log approximation, only valid when the ratio of overtime hours to total hours worked is 

small. In Section IV below, we show that this is the case in the data we examine. 

6

 See Trejo (1991; p.738). 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 6

III. Kinoshitaís Hedonic Wage Curve and the Structure of Compensation with Overtime Pay 

Regulation in the Fixed-Job Model

The data that Trejo used in his study was Current Population Survey data from the midseventies; applying Trejoís empirical strategy to recent data is more difficult because the 

proportion employed in hourly jobs and exempt from FLSA overtime requirements has declined 

considerably since the seventies. The U.S. Department of Labor estimated that the proportion of 

workers exempt from by the FLSA overtime pay provisions declined from 34% of all nonsupervisory civilian workers in 1970 to 16% in 1989, with the largest contractions occurring in 

retail trade, services and local government.7

 Further reductions in workers exempt from FLSA 

coverage were likely to have occurred in retail trade and services establishments since 1989. In 

these industries, establishments below a nominal sales threshold ($500,000 annually) have been 

exempt from both FLSA minimum wage and overtime requirements, but inflation has 

substantially eroded the scope of this exemption since 1989. Also, while substantial exempt 

employment remains in some transportation occupations, much employment in this jobs involve 

long regularly scheduled work hours dictated by the production technologyónot intermittent 

overtime work. For example, in the March 2004 data I used for this study, 51 per cent of those 

who worked as Captains, Mates, and Pilots of Water Vessels (SOC 535021) had usual work 

schedules over 40 hours a week; if crossing the ocean, as many such workers do, itís hard to work 

9 to 5. 

Given the reduced size of the hourly paid FLSA exempt workers in recent data I take a 

somewhat different approach to the empirical analysis. I examine variation in hours and earnings 

among jobs covered by an overtime pay requirement, with comparisons to a ìcontrolî group not 

subject to the overtime pay requirement. To relate these findings to the predictions of the FixedJob Model, I draw on some further development of the Lewis-Barzel analysis by Kinoshita 

(1987), in which the cross section patterns of total job compensation and hours worked in the data 

reflect a hedonic wage locus.8

 

The analysis underlying the Fixed-Job Model addresses a fundamental question: why do 

hours-earnings ìpackagesî vary across jobs? While differences across workers in the demand for 

leisure and other nonwork uses of time likely contributes to the observed variation, Barzel (1973) 

noted that work schedules can also vary between jobs because idiosyncratic production setup 

costs or other workplace-specific circumstances of technology generate heterogeneity across jobs 

 

7

 See Table 1 in Trejo (2003; p. 538) 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 7

in the productivity of employment of otherwise identical job skills. In particular, the value of 

overtime hours to the firm could vary across establishments, depending on the production 

technology, the distribution of market demand, and the costs of coordination of firm work force 

efforts due to worker leave policies and absenteeism. 

Kinoshita (1987) develops a model of the determination of hours worked that can be 

viewed as an explicit Fixed-Job Model. Kinoshitaís model is an application of the Rosen (1974) 

hedonic market model in which hours worked is the primary job characteristic of interest to both 

firms (because it can influence workplace productivity) and workers (because it influences the 

workerís allocation of time). In a given occupational labor market (i.e., jobs requiring similar 

sets of skills) firms and workers sort themselves out between high total earnings/high hours 

worked employment bargains and low total earnings/hours worked employment bargains. This 

variation in earnings-hours packages generates what Kinoshita terms the ìHedonic Wage 

Curveîó the market equilibrium locus of total compensation and hours worked bargains across 

jobs. Holding other job characteristics constant, the slope of the Hedonic Wage Curve is the gain 

(loss) in total compensation that would result from making an alternative employment bargain in 

the job market that has one more (less) hour of work. 

Consider the overtime hour dimension of the employment bargain and denote the slope of 

the Hedonic Wage Curve in the overtime hour region as the Job Market Overtime Premium. 

Figure 1 illustrates how the Job Market Overtime Premium relates to coverage of the FLSA legal 

overtime premium for a job with given productive characteristics. Assuming that total job 

compensation is solely hours-related earnings, the Hedonic Wage Curve is depicted by the thick 

line labeled Φ(H).9

 At point A, with a 40 hour work schedule and no overtime, firms offering this 

employment bargain have valuations of work schedules represented by the curve FA(H) while 

workers choosing this employment bargain have preferences represented by the curve UA(H); 

both curves are tangent to Φ(H) at A. The straight-time wage rate associated with this 

employment bargain, WA , is indicated by the angle of a ray from the origin through A. Note that 

at point A (as well as at point B) there is no necessarily relationship between the straight-time 

wage, (or the overtime wage rate at B), with workerís rates of substitution between income and 

 8

 In his analysis, Trejo also views the Fixed-Job Model as interpreting cross section hours-earnings data as 

an hedonic wage locus, but since he compared jobs with the same hours of work, issues of functional form 

of the hedonic wage locus did not have to be considered. 

9 Φ(H) is constructed as a straight line for simplicity of exposition. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 8

leisure or (implicitly) the firmís rate of substitution between hours and workers10. As in the logic 

of Trejoís use of the Fixed-Job Model, employment bargains with overtime such as illustrated 

point B on Φ(H) may require some downward adjustment in the straight-time wage rate even if 

all jobs offering overtime also required payment of an overtime premium. 

If overtime pay regulation does change the structure, but not the level, of job 

compensation when the Hedonic Wage Curve defines the job market equilibrium, then the 

Hedonic Wage Curve also defines how the structure of compensation is influenced by differences 

in overtime hours between jobs. This point can be illustrated by the following example of the 

structure of job compensation somewhat more general than depicted in Figure 1. Suppose that all 

compensation is cash earnings, and cash earnings for all jobs has the following structure: (1) 

ìstraight-time earningsî for standard hours, S, at the straight time wage rate w; (2) ìovertime 

payî for overtime hours, O, regulated to be 1.5 times w; (3) an annual per-worker payment (a 

ìbonusî), B, which is independent of hours worked. If the Hedonic Wage Curve does define the 

job market equilibrium, then the relationship between total cash earnings, straight-time wages, the 

annual per-worker payment, and the ratio of overtime hours to the standard work week, (O/S), 

across jobs is given by: 

(1) 



 



 



 



 + 



 



 



 



 Φ > = ⋅ +

S

O

S

O (H; H S) w S 1 1.5 1 β

where the β parameter reflects the ratio of hours-related earnings (straight-time and overtime pay) 

to annual per-worker payments in the bonus B. 

Equation (1) determines the combinations of w and B that are consistent with overtime 

pay regulation and the cross section earnings-hours worked locus being located on the overtime 

region of the Hedonic Wage Curve. If (1) holds, it also constrains what adjustments in w and B 

can occur with variation in overtime hours. Taking logs of both sides and using approximations 

appropriate for when the ratio (O/S) is small, the elasticity of the compensation with hours 

worked across jobs in the overtime hour region has the following decomposition: 

(2) ( ) ( ) ( ) 1.5

/ /

ln

/

ln

ln

ln ≈ + +Φ = Φ

d O S

d

d O S

d w

d O S

d

d H

d β . 

 

10 Cutler and Madrian (1998) analyze firm decision-making on the mix of hours worked and employment 

levels, using the Hedonic Wage Curve as the relevant job market constraint on firm choices. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 9

Thus, if the elasticity of total earnings with respect to overtime hours, the first term in (2), is less 

than 1.5 then job market equilibrium can be maintained by either reduction in the straight-time 

wage w and/or the relative size of a bonus supplement β. 

IV. Establishment Data on Job Compensation Costs and Overtime Hours

The research in this paper uses data from a March 2004 cross section of the BLS National 

Compensation Survey (NCS). The NCS provides a representative sample of U.S. private industry 

jobs that allows quarterly estimation of the BLS Employment Cost Index (ECI) and the Employer 

Costs for Employee Compensation (ECEC) series. The NCS sample is selected using a three 

stage stratified design with selection probabilities proportional to employment at each stage; the 

first stage samples areas, the second stage samples establishments, and the third stage samples 

jobs (rather than individual workers in jobsóthe data I use are job averages). At the initiation of 

data collection a BLS field agent visits the sampled establishment to draw a sample of jobs within 

the establishment, and collects information on a number of characteristics of the sampled 

establishment (e.g., industry, establishment employment size) and sampled jobs (e.g., part 

time/full time status, whether covered by a union contract). 

I used an extract of job level micro data from the NCS that was used to estimate the 

ECEC for March 2004.11 The objective of the ECEC statistical program is to estimate the level 

of total employer costs for employee compensation per hour worked for various job market 

aggregates by both occupation and industry (such as ìmanagerial occupationsî, or ìtransportation 

industriesî). Thus, a great deal of effort is directed at initiation of NCS data collection to obtain 

the usual annual work schedule as well as usual annual overtime and annual leave hours (e.g., 

holidays, vacations, sick leave) that are offered in the sampled job. Detailed information on the 

various components of total employer cost is also obtained for sampled jobs. Data collection on 

the components of annual employer costs include: straight time wage rates, supplemental cash 

pay (overtime earnings, shift differentials, bonuses), employer contributions to worker retirement 

and health insurance plans, and payroll taxes such as employer Social Security taxes. 

I restricted the analysis to my estimate of those jobs paid by the hour (an ìhourly jobî), 

jobs that have the potential of being paid for overtime work. NCS data collection includes an 

 

11 To see the latest ECEC statistics, go to https://www.bls.gov/news.release/pdf/ecec.pdf . The micro data 

are collected under a pledge of confidentiality and hence have very restricted access. Researchers can 

submit a research proposal for obtaining access to the data at the BLS office in Washington, D.C.; see 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 10

indicator of whether there the firm has an explicit provision for compensating overtime work, 

whether or not there are overtime hours actually worked; this provision is my indicator for an 

ìhourly jobî. Table 1 shows the estimated incidence of such ìhourly jobsî by major occupation 

group in March 2004. While covering about 80 percent of U.S. private industry jobs and almost 

all jobs in the machine operator, assemblers, and inspector occupations, Table 1 also shows 

relatively few jobs are covered in the executive, administration, and managerial occupations, and 

that coverage is also thin in other occupational groups where salaried compensation is the rule. 

However, in salaried jobs employer measures of hours worked may underestimate hours worked, 

as employers may have little incentive to directly measure hours worked beyond the usual work 

schedule.12

To separate out job costs that could be directly affected by overtime hours from other job 

costs that would be largely independent of overtime hours worked, I followed a taxonomy 

suggested by Ehrenberg and Smith (2000) to construct measures of ìhour-related labor costs for 

job compensationî and ìworker-related labor costs for job compensationî. Exhibit I shows the 

ECEC compensation elements I have categorized in each aggregate cost measure. Since the data 

used are averages over worker values in the job, this classification is not correct in all cases. For 

example, Federal and State Unemployment Insurance payroll taxes are classified as workerrelated labor costs because these taxes are levied upon only the first $7,000 of annual earnings in 

most states, below the annual earnings of most workers. However, for workers with very low 

earnings, these costs would be better considered a component of hours-related labor costs. To 

mitigate problems of misclassification due to low annual earnings, the analysis was restricted to 

jobs having year-round work schedules, thus eliminating seasonal jobs and jobs on academic year 

work cycles from the analysis. 

Since the Fixed-Job Model predicts that the incidence of overtime work can be lower in 

jobs that pay at or ìclose toî the minimum wage if subject to the overtime pay requirements, I 

used the straight-time wage rate information in the sample to code jobs as ìlow wageí jobs if the 

average straight-time wage rate was less than $6 per hour. In many low wage retail and service 

jobs, workers who start at the minimum wage do get an increase if they stay on the job for some 

specified length of time. Thus, a job with an average wage rate above $5.15 per hour can 

 

information at https://www.bls.gov/bls/blsresda.htm . If a researcher does gain access to replicate this 

research, I will be happy to provide my data extracts, codebooks, and logs. 

12 Allegretto, Corcoran, and Mishel (2004) show that the weekly work schedule hours for school teachers in 

the NCS data are significantly lower than usual weekly hours of work of teachers as measured in the 

Current Population Survey (CPS). 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 11

nevertheless include many workers who earn at the minimum and thus subject to the constraint on 

the offering of overtime work emphasized in the Fixed-Job Model. 

I further restricted the sample to compare a set of jobs assumed to subject to the FLSA 

overtime pay regulations (a ìtreatmentî group) with a set of jobs assumed to be not subject to the 

FLSA overtime pay regulations but that could nevertheless require overtime work (a ìcontrolî 

group). The NCS does not ask employers whether sampled jobs are subject or exempt from FLSA 

overtime pay regulations, so the assignment of jobs to a treatment or control group was based on 

the weekly work schedule information provided for the job in data collection. Any hourly job 

with a work schedule of exactly 40 hours a week was assumed to be subject to the overtime pay 

regulations. The NCS data indicate that relatively few full-time hourly jobs (those that 

compensate overtime work) have weekly work schedules over 40 hours a weekósuch jobs 

comprise only 0.57 percent of all employment in hourly jobsóand many of those jobs that do 

have worker weekly work schedules over 40 hours a week are in transportation occupations that 

can be exempt from the FLSA overtime pay regulations.13

The set of hourly jobs for the ìcontrol groupî were those year round part time jobs with 

work schedules of 32 hours a week or less. Of course, sufficiently high overtime hours worked in 

these jobs could push total hours worked over the 40 hour a week threshold and thus trigger 

required time and a half pay.14 However, the distribution of annual overtime hours observed in the 

data for part time jobs indicate that such potential impacts would be minimal. For example, the 

95th percentile for annual overtime hours in the set of jobs with reduced work schedules is about 

166 hours. Suppose that this overtime work was concentrated in 20 weeks a year, requiring 8.3 

hours of overtime per week.15 In this case, a job having a 32 hour week regular schedule would 

require time and a half pay for a total of only 6 of the 166 overtime hours per year. 

Table 2 provides some summary characteristics on compensation and overtime hours for 

the two groups of jobs. As expected, the data indicate that, on average, part time jobs have lower 

wages, have less worker-related costs (i.e., provide less employee benefits), and are less likely to 

require overtime. (That part-time job status appears to be a compensation-determining 

 

13 In addition to the case of Captains, Mates and Pilots of Water Vessels mentioned above, about 23 percent 

of those employment in the occupation Truck DriversóHeavy and Tractor Trailer (long haul trucking) and 

39 percent of employment in the occupation Sailors and Marine Oilers have work schedules over 40 hours. 

These cases are excluded from the analysis. 

14 Of course, employers may negotiate voluntary overtime premiums with their part-time workers, but 

information on such overtime pay is voluntary or mandatory is not collected in the NCS. 

15 Unfortunately, the NCS does not collect data on the distribution of overtime hours worked over the year. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 12

characteristic in of itself will be considered in the design of the statistical analysis.) Among jobs 

with usual overtime hours, the 40 hour a week group of jobs does have a substantially higher 

mean number of annual overtime hours (about 126 hours) than in the reduced hour job group 

(about 46 hours). However, the analysis will focus on comparing variation with overtime hours 

within each group, so it is important to note that the variances in the ratio of annual overtime 

hours to annual work schedule hours in each group of jobs are similar, and we conduct analyses 

using this ratio variable. 

V. Estimation Methodology and Statistical ResultsóEffects on the Incidence of Overtime Work

To analyze the determinants of the incidence of overtime work, I assume that there is a 

cross section function relating expected profits of using overtime to its various determinants, such 

as firm technology, market demand, and the cost of coordinating employment within the firm. 

Thus, given that profits of overtime use have a stochastic component, the firm and job 

characteristics that are expected to influence the profitability of overtime use are also assumed to 

influence the probability of overtime use. 

Given this framework, two propositions follow. First, in the Fixed-Wage Model overtime 

pay regulation introduces a kink in the labor cost curve that is not present in other jobs; for the 

same straight-time wage rate, marginal labor costs are higher and profits are lower with overtime 

use. Thus, holding other job characteristics the same, a given wage rate should be associated with 

a lower probability of overtime in a job with overtime pay regulation than in a job not subject to 

overtime pay regulation. Second, in the Fixed-Job Model jobs paying ìclose toî the minimum 

wage will have lower profits with overtime use than other jobs, because the minimum wage 

prevents downward adjustment of the wage rate to facilitate the optimal employment bargain. 

Hence, the probability of overtime use in a job should be lower in low wage jobs (all other things 

being equal) when subject to overtime pay regulation. 

To test these propositions, I estimate two variants of a cross section probability of 

overtime function, allowing the effects of the straight-time wage rate and being a low wage job to 

vary whether a job has a 40 hour a week work schedule (presumed covered by overtime pay 

regulation) or a work schedule of 32 hours or less a week (presumed not to be impacted by 

overtime pay regulation). Among the variables obtained in NCS data collection that are used to 

control for the effects of other factors on overtime use in these probability functions are the 

industry of the establishment, firm establishment employment, whether leave is provided workers 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 13

on the job, whether the job is a union job, and the level of worker-related labor costs. Providing 

leave to workers could increase the costs of coordination of the labor force due to absenteeism 

and thus increase the probability of overtime use. Higher levels of worker-related labor costs 

could induce firms to use the existing labor force more intensively, increasing the probability of 

overtime use. 

The two variants used for estimating overtime use incidence equations were a linear 

probability function, estimated by OLS, and the odds ratio formulation of the logit function. The 

odds ratio formulation of the logit was used to facilitate comparisons to the results for linear 

probability function, exploiting the property of the logit functional form that the marginal impact 

of a variable on the odds of overtime use does not depend on the level of that variable. Thus, 

testing the null hypothesis for a zero coefficient on a variable in a linear probability model is 

similar in intent to testing the null hypothesis for a coefficient of one on that variable in the odds 

ratio formulation of the logit function. Looking at the broad pattern of results in this way, the 

results of the two estimations are quite consistent; that is, variables having a negative coefficient 

in the estimated linear probability function also had a coefficient of less than one in the odds ratio 

formulation of the logit function. The same set of statistical controls were used in both 

estimations, except that the linear probability function included very detailed industry coding of 

jobs (equivalent to 777 fixed effects) while the logit function used more aggregated industry 

coding. The sample size for both estimates was 20,835 jobs. 

The results of the analysis of incidence are given in Table 3. The estimated impact of 

wages rates in both probability functions is consistent with the Fixed-Wage Model predictions. 

The Fixed-Wage Model predicts the impact of a higher wage rate should be greater in jobs 

subject to overtime pay regulation. Among jobs with 40 hour work week schedules, both 

estimates indicate a statistically significant reduction in the probability of overtime with an 

increase in the straight-time wage. (In the estimated logit function the coefficient on the straighttime wage rate, 0.3998, indicates that the odds fall with an increase in the wage.) In contrast, the 

estimated impacts of a higher wage rate, while also negative, are smaller in magnitude among 

jobs with work schedules of 32 hours or less, and are not statistically significant. The evidence for 

the effects predicted by the Fixed-Job Model from a job having a wage close to the minimum 

wage is not as strong. Being a low wage job does reduce the probability of overtime use among 

jobs with 40 hour work schedules in both of the estimated probability functions, but neither is 

statistically significant. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 14

Table 3 also reports the estimated effects of other selected job and establishment 

characteristics in the probability functions. As indicated by the summary comparisons in Table 2, 

part-time jobs have a much lower probability of overtime use, holding other job and 

establishment characteristics constant. Higher worker-related labor costs raise the probability of 

overtime use, in line with the expectation that labor would be used more intensively given greater 

worker-related costs. The probability of overtime use rises with establishment size and whether 

work leave is provided on the job. Also, the results indicate a small but statistically significant 

lower probability of use of overtime among union jobs.

VI. Estimation Methodology and Statistical ResultsóEffects on the Level and Structure of 

Compensation

In this section we examine data that pertain to the effects of overtime pay regulation on 

job compensation among the subset of jobs that do require overtime. The general strategy is to 

assess whether the predictions from the Fixed-Job Model or the Fixed-Wage Model are more 

consistent with cross section equations estimated by Ordinary Least Squares (OLS) that relate the 

level and structure of compensation to overtime hours worked, comparing jobs with 40 hour work 

schedules with jobs with reduced weekly work schedules. 

Two alternative specifications of the compensation equations are estimated. First, the log 

compensation measuresótotal hours-related labor costs, straight-time wage rates, and total 

worker-related labor costs--are each specified to be a linear function of the ratio of annual 

overtime hours to the annual work schedule, allowing the coefficients on these variables to differ 

between part time jobs and forty hour a week jobs. The estimations using this specification 

provide a first approximation to whether higher overtime hours induce changes in the structure of 

job compensation that are suggested by the Fixed-Job Model and shown in equation (2) above. 

The second specification is more general, allowing the log compensation measures to be a 

quadratic function of overtime hours (again, allowing the coefficients on these variables to differ 

between part time jobs and forty hour a week jobs). Using the quadratic specification, levels of 

hour-related labor costs expected under the Fixed-Wage Model are compared to the levels of 

hour-related costs predicted by the cross section relationships, holding job characteristics 

constant. 

The use of a ratio of overtime to scheduled work hours in the first specification is 

motivated from a log decomposition of the effect of the various components of hours worked on a 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 15

log compensation measure. Defining net hours worked as H = S + O ñ L (Hónet hours worked; 

Sóusual work schedule hours; Oóovertime hours; Lóleave hours), then when O/S and L/S are 

small, ln (1 + O/S) ~ O/S, ln (1 ñ L/S) ~ -L/S, and ln H ~ ln S + O/S ñ L/S. Assuming that these 

different margins for hours worked can have different effects on compensation I estimated the 

following relationship: 

(3) i

i

i

i

i

P

i

i

i F

M

i ij S

L

S

O

S

O C α βX δ S κ  + ε 

 



  +



 



  + Φ 

 



 ln = +∑ + ln + Φ

1

. 

In (3), Ci represents one of the three compensation measures and εi is an error term. The Xijs in 

the equation include those wage-determining job and establishment characteristics obtained in 

NCS data collection I assume are exogenous variables. Even though the specific hours worked 

package is subject to bargaining with workers and thus endogenous, I assume that the firm 

unilaterally decides whether to create a part-time job or a full-time job on the basis of skill 

set/productivity decisions about job use in the firmís production technology.16 (For example, 

because of the time limitations, part time workers in a given occupation may have fewer job 

responsibilities.) Thus, the Xs include a set of detailed (6-digit Standard Occupational 

Classification) occupation / full time / part time combinations as dummy variables, representing a 

total of 769 fixed effects, so that the specification in (3) allows for a separate part time / full time 

compensation differential in each detailed occupation. Other job characteristics included in the 

Xs include whether the job is a union job or whether incentive pay is used. The Xs also include 

several establishment characteristics: dummy variables for major industry classification of the 

establishment, the Census region location of the establishment, and whether the establishment is 

located outside a metropolitan area, and the log of establishment employment as an indicator of 

establishment size. 

 The remaining terms in equation (3) represent the various effects on the 

compensation measures with determination of the hours worked package for the job. These 

variables are endogenous and must be uncorrelated with the error term for OLS estimation to 

obtain unbiased estimates of the hours worked parameters. (We discuss this assumption in more 

detail below.) The coefficient on the log of annual work schedule hours, δ, reflects the effect of 

variation in weekly work schedule hours only among part time jobs, since annual work hours is 

the same among all the jobs with 40 hour work schedules. The coefficient ΦF measures the 

impact of variation in a compensation measure from the overtime ratio among forty hour work 

 

16 Using NCS data, Lettau (1997) shows that there is a full time/part time compensation differential among 

identical occupations within the same establishment. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 16

schedule jobs and the coefficient ΦP measures the impact among reduced work hour schedule (32 

hours or less per week) jobs; I assume the effects of overtime pay regulation are minimal in the 

latter group. Finally, the coefficient on the leave ratio, κ, should be negative (more leave implies 

fewer hours worked) if representing tradeoffs on the leave dimension of the hours package along 

the Hedonic Wage Curve. 

Both the Fixed-Wage Model and the Fixed-Job Model suggest systematic differences 

between ΦF and ΦP. In the equation for log total hours-related compensation, ΦF and ΦP provide 

estimates of the elasticity of compensation with respect to overtime hours. In the Fixed-Wage 

Model ΦF would be determined by the statutory overtime premium and thus should be close to 

(but below) 1.5 with positive overtime hours.17 In contrast, in the Fixed-Job Model, the elasticity 

is determined by the Job Market Overtime Premium, which would depend on the curvature of the 

Hedonic Wage Curve in the relevant overtime region. In the equation for log straight-time wage, 

ΦF and ΦP provide estimates of the adjustments to straight-time wage with varying overtime use. 

As indicated by equation (2) above, the Fixed-Job Model suggests that if the Job Market 

Overtime Premium is less than 1.5, then in full-time jobs, the straight-time wage could fall with 

increased overtime use so that ΦF is expected to be negative in the straight-time wage equation. 

Also, in the Fixed-Job Model Equation, equation (2) above also suggests that among forty hour a 

week jobs an alternative or additional way for the structure of compensation jobs to adjust to the 

constraint of payment of the FLSA overtime pay requirement would be for total worker-related 

costs for compensation to fall with higher overtime; in the Fixed-Job Model ΦF could be negative 

in the equation for worker-related costs for compensation as well. 

I obtain OLS estimates of the equations, accounting for correlation of errors across jobs 

in the same establishment. To be able to use these results to evaluate the predictions of the FixedJob Model I have to assume that the relationships estimated allow some inferences about the Job 

Market Overtime Premiumóthe implicit price of an hour of overtime along the relevant region of 

the Hedonic Wage Curve. However, Hwang, Reed, and Hubbard (1992) show that OLS 

estimation of a hedonic wage locus from cross section data can produce biased estimates of the 

implicit price of a job characteristic when there is a strong correlation between an unmeasured 

component of job productivity and worker demand for that job characteristic. When hours worked 

is the relevant job characteristic, workers with more valuable labor market skills may choose to 

increase their leisure through taking jobs with fewer scheduled hours, more leave hours, or 

requiring less overtime hours, assuming the demand for leisure is a normal good among workers 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 17

generally. When overtime hours is a channel for worker demand for leisure and that the statistical 

design does leave unmeasured many determinants of worker skills, OLS estimation could 

generate downward-biased estimates of the effect of overtime hours on the compensation 

measures. Also, the lack of data collection on worker characteristics in the NCS prevents the use 

of instrumental variable estimation (that is, using instrumental variables for the hours worked 

measures that are correlated with worker preferences for leisure but uncorrelated with the value of 

worker market skills, such as whether the worker is a multiple job holder). 

However, there are several characteristics of the data I use that can limit the degree of 

OLS bias in estimating the relationship between overtime hours and job compensation. First, the 

set of exogenous variables account for a high proportion of the variance in the measures of job 

compensation; adjusted R-squares are about 0.80 in the equations for total hour-related and 

worker related job compensation and about 0.73 in the equation for the straight-time wage rate. 

Hwang, Reed, and Hubbard (1992) show that as the R-square of an OLS estimated regression 

approaches 1, then OLS bias approaches zero. Secondly, a data analysis I completed for an 

earlier paper (Barkume (2006)) suggests that OLS bias is likely to be a problem for leave hours, 

but not for overtime hours or usual work schedule hours.18

Table 4 shows the results of estimating the parameters for the three compensation 

measures using the specification shown in (3) above. The first two rows report the coefficients 

for ΦF and ΦP in the three compensation equations. These results suggest some support for the 

Fixed-Job Model: (1) the estimated elasticity of hours-related labor costs for compensation, while 

greater than 1.0, is substantially below 1.5 in jobs with forty hour work schedules and not 

statistically different from the corresponding elasticity in the reduced-hour jobs; (2) In the 

straight-time wage equation ΦF is negative and statistically significant while the ΦP coefficient is 

not statistically significant. But note that in the worker-related costs equation, the effects of 

overtime hours are not statistically significant in forty hour jobs, which is inconsistent with the 

proposition that worker-related costs are a channel for adjusting total labor costs under an 

overtime pay constraint. 

 17 In the Fixed-Wage Model, the average wage rises with more overtime, while the marginal overtime wage 

remains constant at 1.5 times the straight-time wage. 

18 In the earlier paper, I estimated residuals from a regression of log per-hour compensation measures on 

the exogenous variables but excluding the hours worked measures. I then regressed these residuals on the 

measures of the three dimensions of hours worked. I did find a significant negative impact of leave hours 

on the compensation residuals (see Barkume (2006; Table 4)). 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 18

Some of the other results of the estimation shown in Table 4, while not directly bearing 

on evidence for the Fixed-Job and Fixed-Wage Models, are of interest in themselves: (1) Whether 

a job is in the comparison group (32 hour or less work schedule) does not have statistically 

significant effects on any of the compensation variables, reflecting that the effect of part-time job 

status has been captured with the occupational-specific part-time/full time differentials in the 

equation specification; (2) The coefficients on the leave ratio in all three equations are consistent 

with the findings of Altonji and Usui (2005), who found that more leave is associated with higher 

compensation, but also inconsistent with the estimates reflecting a leave hour segment of the 

Hedonic Wage Curve; (3) Because the coefficients for the employment size, union job, incentive 

pay, and non-metropolitan job location variables are almost identical in the hours-related total 

compensation and straight-time wage rate equations these job characteristics affect hours-related 

labor compensation only through determination of wage rates, not through determination of hours 

of work. 

To compare predicted labor cost differentials between jobs with different amounts of 

overtime hours to those expected under the Fixed-Wage Model, the labor cost equations were reestimated using a more flexible functional form. Specifically, in the regression equations for the 

log compensation measures specified in equation (3) above, the terms ΦF (Oi/Si) + ΦP (Oi/Si) were 

replaced with a quadratic set of terms having the form: ηF Oi+ ηP Oi + λF (Oi)

2

+ λF (Oi)

2 , again 

allowing the coefficients to vary between the 40 hour full time jobs and the reduced hour jobs.19

Using estimates of the ηF and λF coefficients, predicted log differentials for total hours-related 

labor costs and the straight time wage rate were generated for selected points in the distribution 

of annual overtime hours among the 40-hour jobs, relative to jobs without overtime with the same 

job characteristics. As was the case for the results using the overtime ratio specification, shown 

in Table 4, overtime hours had no statistically significant impacts on the log of worker-related 

labor costs in the quadratic specification. 

Table 5 summarizes the results of this exercise. In the first column are the points of the 

distribution of overtime hours among 40 hour jobs to be compared: 25th percentile (29.7 hours), 

median (82.8 hours), 75th percentile (175.3 hours), and the 90th percentile (281.6 hours). In the 

second column is the frame of reference for the predictions from the estimates, the expected log 

hours-related labor cost differential using the simplest Fixed-Wage Model (straight-time wage 

rate does not vary with overtime hours)--ln [(1.5 overtime hours)/annual work schedule hours]. In 

 

19 It turns out that the differences between the quadratic coefficients for the part time jobs and those for the 

full time jobs were not statistically significant. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 19

the third column is the predicted log hours-related labor cost differential, holding other job 

characteristics constant. Comparing the second and third columns, the predicted log differential in 

hours-related labor costs is about 85 % of the log differential expected with the simplest FixedWage Model. The third column gives the predicted adjustment in the straight-time wage with 

increased overtime. (Because the straight-time wage rate equation is independently estimated, the 

log differentials in the second and third column columns do not exactly sum to the log differential 

expected with the Fixed-Wage Model.) Increased overtime levels are associated with downward 

adjustments in the straight-time wage rate but these predicted adjustments are modest.20 Given the 

range of observed overtime hour use, only small reductions in the straight-time wage rate are 

consistent with the labor cost differentials falling significantly short of those predicted by the 

Fixed-Wage Model. For example, the data in Table 5 show that a job having annual overtime 

hours at the 90th percentile of the distribution, about 282 hours per year, the log wage differential 

predicted relative to a job without overtime would be only about -0.03 for the log cost differential 

to be 85% of that predicted by the Fixed-Wage Model. 

VII. Conclusions and Extensions

The Fixed-Job Model implies that U.S. overtime pay regulation should have little effect 

on the incidence of overtime and the level of job compensation, except for legal barriers to 

worker-firm transactions such as the minimum wage. I did find that increases in labor costs 

induced by more overtime work fall substantially short of the increases predicted by the FixedWage Model, and that this mitigation of cost growth is in part due to modest declines in the 

straight-time wage with more overtime work. However, it is more difficult to establish that these 

results reflect the importance of the Fixed-Job Model in the operation of the labor market. My 

estimation of the labor costs ñ hours worked relationships does not rule out the possibility that 

workers with more overtime hours are less productive in ways not captured by the statistical 

controls in the analysis, with those workers commanding higher compensation also avoiding jobs 

with more overtime. Furthermore, the results of the incidence analysis are consistent with the 

predictions of the Fixed-Wage Model. Higher wages reduced the probability of offering overtime 

in full time jobs subject to overtime pay regulation but not in part time jobs not requiring 

overtime pay premiums, consistent with the kink in the labor cost structure with overtime hours 

implied by the Fixed-Wage Model. Also, I found that full time jobs subject to overtime pay 

regulation but paying at or close to the minimum wage did not have a lower probability of 

overtime work, as would be predicted by the Fixed-Job Model. 

 

20 In the straight-time wage rate equation, the hours squared term is not statistically significant. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 20

A more clear-cut test of the relevance of the Fixed-Wage and Fixed-Job Models than in 

analysis of cross section data may be found in examining the effects of a recent change in the 

U.S. overtime pay regulation known as the ìFair Pay Initiativeî21. The Fair Pay Initiative 

regulations, introduced in August 2004, revised somewhat the rules for exemption of the ìwhite 

collarî (executive, administrative, and professional) jobs. Any white collar job paying a salary of 

less than $455 per week is now covered by the overtime pay requirement without exception. Any 

white collar job previously determined to require overtime by virtue of job duties is nonetheless 

now exempt from the overtime requirement if worker earnings in the job exceed $100,000 per 

year. For white collar workers within those earnings bounds, there are revisions in the job tests 

for exemptions to overtime requirements affecting certain occupations, such as low-level salaried 

supervisors. The Fixed-Wage Model should predict earnings increases in lowly paid salary jobs 

requiring overtime and earnings decreases in salaried jobs having overtime pay protection but 

earnings over $100,000 per year; in contrast, the Fixed-Job Model predicts no systematic earnings 

differences should occur with these regulatory changes. 

The Fixed Job Model does place considerable emphasis on the ability of workers and 

firms to contract the optimal employment bargain. But it may provide a useful supplement to 

studying other labor market issues, instead of just assuming that firms and workers always act as 

price-takers in the job market. For example, Barzel (1973; pp. 230-233) suggested that a possibly 

viable response of workers and firms to a minimum wage increase would be an increase in hours 

per worker in the employment bargain (lowering average hourly earnings); if so, the impact of a 

minimum wage would be reduced employment and increased hours per worker. An overtime pay 

requirement would reduce the potential for revising the employment bargain in this way, so that 

this potential effect of minimum wage increases on hours worked suggested by the Fixed-Job 

Model would be expected to be concentrated in part time jobs. 

 

21 For background information, see the relevant Department of Labor website page 

http://www.dol.gov/esa/regs/compliance/whd/fairpay/main.htm . For an early analysis of the consequences 

of this regulatory initiative (using the Fixed-Wage Model!) see Eisenbrey (2004). 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 21

References

Allegretto, Sylvia A., Sean P. Corcoran, and Lawrence Mishel, How Does Teacher Pay 

Compare? Methodological Challenges and Answers. Economic Policy Institute, 2004. 

Altonji, Joseph G. and Emiko Usui, ìWork Hours, Wages, and Vacation Leave,î NBER Working 

Paper No. 11693 (October 2005). 

Barkume, Anthony J., ìDoes Average Hourly Earnings Indicate the Price of Labor? Some Cross 

Section Evidence for U.S. Private Industry,î paper presented at the January 2006 

Meetings of the Allied Social Sciences Association, Boston Masssachusetts. 

Barzel, Yoram, ìThe Determination of Daily Hours and Wages,î The Quarterly Journal of 

Economics, Vol. 87, no. 2 (May 1973), 220-238. 

Costa, Dora L. ìHours of Work and the Fair Labor Standards Act: A Study of Retail and 

Wholesale Trade, 1938-1950,î Industrial and Labor Relations Review, vol. 53, no. 4 

(July 2000), pp. 648-64. 

Cutler, David M. and Madrian, Brigitte. ìLabor Market Responses to Rising Health Insurance 

Costs: Evidence on Hours Worked,î The Rand Journal of Economics, Vol. 29, No. 3. 

(Autumn, 1998), pp. 509-530. 

Ehrenberg, Ronald G., and Robert S. Smith, Modern Labor Economics: Theory and Policy

(Seventh Edition) Addison, Wesley Longman, 2000. 

Eisenbrey, Ross, ìLonger Hours, Less Payî Briefing Paper, Economic Policy Institute, July 2004. 

Hwang, Hae-shin; Reed, W. Robert, and Hubbard, Carlton. ìCompensating Wage Differentials 

and Unobserved Productivity,î The Journal of Political Economy, vol. 100, No. 4 (Aug. 

1992), pp. 835-858. 

Kinoshita, Tomio, ìWorking Hours and Hedonic Wages in the Market Equilibrium,î The 

Journal of Political Economy, Vol. 95, no.6 (December 1987), pp. 1262-1277. 

Lettau, Michael K., ìCompensation in Part-Time Jobs versus Full-Time Jobs: What if the Job is 

the same?î Economic Letters, Vol. 56, (1997), pp. 101-106. 

Lewis, H. Gregg, ìEmployer Interest in Employee Hours of Work,î mimeo, Duke University, 

1969. 

Rosen, Sherwin, ìHedonic Prices and Implicit Markets,î Journal of Political Economy, Vol. 82 

(1974), pp. 34-55. 

Trejo, Steven J., ìThe Effects of Overtime Pay Regulation on Worker Compensation,î American 

Economic Review, vol. 81, no.4 (September 1991), pp. 719-740. 

Trejo, Steven J., ìDoes the Statutory Overtime Premium Discourage Long Workweeks?î 

Industrial and Labor Relations Review, vol. 56, no. 3 (April 2003), pp. 530-551. 
 

Table 1. Estimated employment incidence by occupation of jobs with provisions for 

paying overtime (proxy for ìhourly jobsî in the analysis), U.S. private industry, 

March 2004 

 

Major Occupational Group 

Employment in 

jobs with overtime 

provisions 

(millions) 

Share of total 

occupational 

employment 

Professional and technical 8.994 0.58 

Executive, administrative and managerial 1.705 0.19 

Sales 9.453 0.76 

Administrative support, including clerical 17.300 0.91 

Precision production, craft and repair 9.959 0.93 

Machine operators, assemblers, and inspectors 7.220 0.99 

Transportation and material moving 5.006 0.90 

Handlers, equipment cleaners, helpers and 

laborers 

7.951 0.96 

Service workers , except private household 19.300 0.94 

All US private industry 86.892 0.80 

Source: Job level microdata from BLS National Compensation Survey, March 2004 cross 

section, using current employment weights. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 1

Exhibit I 

ECEC Data Elements Measuring Hour-Related and Worker-Related Labor Costs 

Hour-Related Labor Costs for Job Compensation (primarily a function of hours worked)

Straight-time wage payments 

Overtime costs 

Social Security payroll taxes (employer share) 

Medicare payroll taxes (employer share) 

Workers compensation insurance 

Defined contribution component of employer pension (employer contribution) 

Worker-Related Labor Costs for Job Compensation (largely independent of overtime)

Vacation and Holiday Paid Leave 

Paid Sick Leave 

Other Paid Leave (e.g., for jury duty) 

Payments for Shift Differentials 

ìNonproductionî Bonuses (unrelated to worker hours or output) 

Severance Payments 

Supplemental Unemployment Insurance 

Health Insurance (employer contribution) 

Life Insurance (employer contribution) 

Sickness and Accident Insurance (employer contribution) 

Long Term Disability 

Defined Benefit component of employer pension 

Federal and State Unemployment Insurance payroll taxes 

Note: Labor cost components are data elements in the estimation of the BLS Employer 

Costs for Employment Cost (ECEC), as part of the National Compensation Survey. All 

labor costs are employer contributions. Classification of ìearnings-relatedî and ìworkerrelatedî costs follows Ehrenberg and Smith (2000). 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 2

Table 2. Summary data on overtime use and job compensation 

Hourly jobs with 40 hour 

per week work schedules 

Hourly jobs with work 

schedules 32 hours or less 

Sample size (no. of jobs) 16,219 4,622 

Estimate of employment 

coverage (millions) 

48.36 20.86 

Straight-time wage (dollars 

per hour) 

14.95 

(7.51) 

9.52 

(7.04) 

Ratio, worker-related costs 

to hour-related costs 

0.249 

(0.151) 

0.109 

(0.154) 

Fraction of employment in 

low wage jobs (wage below 

$6 per hour) 

0.018 0.172 

Fraction of employment in 

jobs using overtime 

0.767 0.281 

Jobs with overtime hours: 

 -Usual annual hours of 

overtime work 

126.2 

(148.2) 

45.6 

(90.87) 

 -Ratio, annual overtime 

hours to annual work 

schedule hours 

0.0607 

(0.0712) 

0.0407 

(0.0842) 

Source: Job level microdata from BLS National Compensation Survey, March 2004 cross 

section, using current employment weights. Data in parenthesis are standard deviations. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 3

Linear Probability Model Logit specification 

(Coefficients are odds ratios) 

Straight-time wage rate (log): 

óin jobs with 40 hour work 

schedules 

-0.1584* 

(0.0250) 

0.3988# 

(0.0636) 

óin jobs 32 hour or less work 

schedules 

-0.0293 

(0.0340) 

0.9632 

(0.1914) 

Reject hypothesis that 

coefficients are equal? YES (t = 3.49) YES (t = 4.14) 

Low wage job (straight-time 

wage less than $6/hr): 

 

óin jobs with 40 hour work 

schedules 

-0.0969 

(0.0629) 

0.4795 

(0.1811) 

óin jobs 32 hour or less work 

schedules 

0.0595 

(0.0445) 

1.8520# 

(0.0544) 

Reject hypothesis that 

coefficients are equal? YES (t = 2.09) YES (t = 2.88) 

Estimated effects of other selected job and establishment characteristics:

--Whether job has work 

schedule of 32 hours or less 

-0.5791* 

(0.0913) 

0.0366# 

(0.0197) 

--Worker-related labor costs 

for job compensation (log) 

0.0900* 

(0.0094) 

1.6869# 

(0.1017) 

--Whether job provides paid 

vacation or sick leave 

0.1563* 

(0.0273) 

2.6669# 

(0.4487) 

--Establishment size (log) 0.0191* 

(0.0056) 

1.1104# 

(0.0316) 

--Whether union job -0.0496* 

(0.0198) 

0.8671# 

(0.1109) 

Adjusted R-square (LPM); 

Pseudo R-square (Logit spec.) 

0.4220 0.2602 

Table 3. Estimated effects of straight-time wage rates on overtime incidence: Job level 

microdata from BLS National Compensation Survey, March 2004 cross section; N=20,835. Dependent 

variable is whether job requires usual overtime work. Linear probability model: least squares estimation, 

allowing for correlation of errors across jobs within an establishment; fixed effects for 6-digit NAICS code 

of establishment (777 industries). Logit estimation: # indicates rejection of the null hypothesis of no 

impact on the odds ratio (95% confidence interval); statistical controls for industry are 2-digit NAICS 

codes. Both specifications also include controls for region and major occupational group of job, whether 

job is located outside an SMSA or uses incentive pay, and the length of the annual work schedule in parttime jobs. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 4

 

Job characteristic: 

Total hour-related 

labor costs for job 

compensation (log)

Straight-time wage rate 

per hour (log) 

Total worker-related 

labor costs for job 

compensation (log) 

overtime hours to 

work schedule ratio: 

 

óin jobs with 40 hour 

work schedules 

1.0521* 

(0.0590) 

-0.1795* 

(0.0608) 

-0.0456 

(0.1557) 

óin jobs with 32 

hours or less work 

schedule 

0.8185* 

(0.1900) 

-0.0870 

(0.2332) 

-0.7218 

(0.4304) 

Other job 

compensation 

variables: 

 

--whether job has 32 

hour or less work 

schedule 

-0.0686 

(0.0484) 

-0.0944 

(0.0522) 

-0.1313 

(0.0941) 

--annual work 

schedule (log) 

0.9156* 

(0.0310) 

-0.0715* 

(0.0324) 

0.7812* 

(0.1118) 

--whether job provides 

any leave hours 

0.0339 

(0.0292) 

0.0375 

(0.0301) 

0.8371* 

(0.0965) 

--ratio of leave hours 

to work schedule, in 

jobs providing leave 

0.6411* 

(0.1442) 

1.495* 

(0.1475) 

10.7557* 

(0.4118) 

--establishment size 

(log) 

0.0297* 

(0.0027) 

0.0296* 

(0.0140) 

0.0831* 

(0.0064) 

--union job 0.2217* 

(0.0140) 

0.2302* 

(0.0140) 

0.4660* 

(0.0284) 

--establishment located 

outside SMSA 

-0.0661* 

(0.0113) 

-0.0621* 

(0.0114) 

-0.0073 

(0.0319) 

--whether job uses 

incentive pay 

0.1136* 

(0.0277) 

0.1181* 

(0.0271) 

0.0232 

(0.0435) 

Adjusted R-square 0.832 0.734 0.800 

Table 4. Estimated effects of overtime hours on job compensation (cross section). Job level 

microdata from BLS National Compensation Survey, March 2004 cross section; N=15,069. All jobs in the 

estimation sample require usual overtime work. Least squares estimation, allowing for correlation of errors 

across jobs within an establishment. Data in parenthesis are standard errors; an asterisk represents rejection 

of a zero coefficient using a 99% confidence interval. Other regressors were detailed occupation/full 

time/part time combinations (equivalent to 769 dummy variables) as well as dummy variables for region 

and industry. 
ìSome New Evidence on Overtime Use, Total Job Compensation, and Wage Ratesî 

 5

Annual overtime 

hours (percentile in 

distribution of annual 

overtime hours, jobs 

with 40 hour weekly 

work schedules): 

Expected log 

differential in hoursrelated labor cost to 

same job without 

overtime, Fixed-Wage 

Model1

Predicted log 

differential in hoursrelated labor cost to 

job without overtime, 

holding other job 

characteristics 

constant 

Predicted log 

differential in 

straight-time wages to 

job without overtime, 

holding other job 

characteristics 

constant 

29.7 (25th percentile) 0.0211 0.0177 

(0.0015) 

-0.0034 

(0.0015) 

82.8 (50th percentile) 0.0579 0.0490 

(0.0041) 

-0.0094 

(0.0040) 

175.3 (75th percentile) 0.1190 0.1019 

(0.0081) 

-0.0193 

(0.0079) 

281.6 (90th percentile) 0.1848 0.1603 

(0.0120) 

-0.0300 

(0.0111) 

Table 5. Predicted hours-related earnings and straight-time wage rate differentials 

in full time jobs with annual overtime hours relative to jobs with no overtime--

quadratic specification of the effects of overtime hours. Standard errors of estimates 

are given in parenthesis. Same data and covariates as results in Table 4, except that the 

ratio of overtime hours to the usual work schedule is replaced with annual overtime hours 

and the square of usual overtime hours, and the estimation allows quadratic coefficients 

to vary if job has a part-time work schedule. 

(1) Ln [(1.5 annual overtime hours)/2080] 
Figure 1. The Hedonic Wage Curve and Payment of Overtime Premiums

φ(HB) 

φ(HA) 

φ(H)

Total Hours-Related Earnings 

Hours Worked

UA(H)

FA(H)

WA * H

 40 

HB

B

A

WB (1.5

H-20) 

WB * 40