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OCR for page 171
I`~ Work Experience,
I Job Segregabon, and Wages
MARY CORCORAN, GREG J. DUNCAN, and
MICHAEL PONZA
Women are a vital part of today's labor
force, and work is clearly an important part
of their lives. Women constituted more than
two-fifths of the labor force in 1978, and
almost 60 percent of all women aged 18 to
64 were employed in 1978. Almost all women
work at some point in their lives, and their
earnings are often necessary to ensure ad-
equate family support. In 1978 nearly two-
thirds of the women working were either
presently unmarried or married to men
earning less than $10,000 per year (in 1977.
There is considerable evidence that men's
and women's work participation patterns
aider with men working continuously after
completing school and women moving in
and out of the labor force to accommodate
family and child-rearing duties. Women earn
considerably less than men do. Since 1930
the median salary of full-time, full-year
women workers has been about 60 percent
of the median salary of men who work full
This paper was supported by a grant from the Alfred
P. Sloan Foundation.
~ These figures are taken from U.S. Department of
Labor, Women's Bureau, 10 Facts on Women Workers,
Washington, D.C., August 1979.
171
time, full year. Women and men also have
very different occupations. Treiman and
Hartmann (1981) show that 70 percent of
the men and 54 percent of the women in
the labor force are concentrated in occu-
pations dominated by their own sex. Unlike
men, women are heavily concentrated in a
few job categories—secretarial work, sales,
teaching, nursing, and various service oc-
cupations.
The most prominent economic explana-
tion linking labor supply patterns and wages
is human capital theory.2 Human capital it-
self is defined as worker skills or qualifica-
tions acquired through schooling or on-the-
job training. An individual worker's stock of
human capital can be increased by the proc-
ess of investment. Investments have an op-
portunity cost (in terms of forgone earnings
as well as of the direct costs of training) and
a return (in the form of higher subsequent
earnings). Human capital theory particu-
2 Human capital theory is quite similar to the func-
tionalist theory of Davis and Moore. It has been argued
that this theory underlies much of the empirical work
in social stratification (see Horan, 1978~.
OCR for page 172
172
MARY CORCORAN, GREG [. DUNCAN, AND MICHAEL PONZA
larly emphasizes investment in formal
schooling (Becker, 1975) and in on-thejob
training (Mincer, 1974~. Workers are pre-
sumed to choose freely among jobs with dif-
ferent amounts of training, and wages are
presumed to reward past investments in ed-
ucation and training in a similar way for all
workers.
In recent years human capital theory has
been expanded to deal with the structure of
female wages. Some of its proponents have
argued that the sex division of labor within
the home generates sex differences in pat-
terns of investment in work-related human
capital and that this in turn generates sex
differences in wages and the sex segregation
of occupations (Mincer and Polachek, 1974;
Polachek, 1976, 1979, 1981; Mincer and Ofek,
19821. These arguments have focused par-
ticularly on sex differences in patterns of
labor force participation.
This paper investigates human capital
theory's predictions about the relationships
between patterns of work, wages, and job
segregation. The paper is in four major parts.
The first summarizes human capital theo-
retical models. In the next section we use
13 years of data from the Pane} StudY of
Income Dynamics (PSID) to describe and
compare men's and women's patterns of work
participation. These comparisons focus on
aspects of lifetime work experience that are
hypothesized to be important for women—
duration of work and nonwork periods, the
extent to which labor market experience in-
volves part-time work, and the sex typing
of experience (i. e., the extent to which past
experience was in female-dominated occu-
pations). The third section reviews past re-
search on the extent to which different as-
pects of work experience influence wages
and sex typing of a person's current lob. ant}
it also includes the results from our own
analyses of these issues based on 13 years
of PSID data. Finally, we discuss the the-
oretical and policy conclusions drawn from
our reviews of past research and from our
own research.
WORK HISTORY, WAGES, AND JOB
SEGREGATION: THEORETICAL MODELS
Work Experience and Earnings
In the human capital model, investments
in on-thejob training are considered to be
critical determinants of wages (see Becker,
1975; Ben-Porath, 1967; Mincer, 1974; and
Rosen, 1972~. On-thejob training has a cost,
since time spent in training is assumed to
be time diverted from production, and pro-
duction presumably determines earnings.
On-thejob training also has a return in the
form of higher later earnings. The following
function describes this hypothetical rela-
tionship:
~—~
F:. = E, + 2, rC' = Ye + C`, (~1)
i=o
where En is earnings capacity in year t, Es
is earnings that would be received in the
absence of any postschoo! training, Ci is the
dollar cost of investments in human capital
in the ith year, Y' is earnings in the tth year,
C' is dollar cost of investments in the tth
year, and r is rate of return to investments
in human capital.
If we assume that total benefits of an in-
vestment increase as the payoff period in-
creases and that the marginal costs of in-
vestments are upwardly sloping in a single
time period, it can be shown that a profile
of investment ratios (CilE`) that are large at
first and then decline over time maximizes
the present value of expected lifetime earn-
ings (see Ben-Porath, 19671. That is, the
proportion of one's earnings capacity in-
vestec3 in on-thejob training will be high in
the early years and then will decline rapidly.
The human capital mode! assumes that
workers freely choose among a variety of
jobs each with a different combination of
training and productive work. It generally
views training and productive work as mu-
tually exclusive activities, and, thus, ac-
cording to the mociel, employers will pay
OCR for page 173
WORK EXPERIENCE, ,~
SEGREGATION AND WAGES
. _
173
less for jobs with training than for similar
jobs that do not provide training. An im-
plication of the mode} is that wages grow
with experience because workers are ac-
quiring additional subs as they increase their
experience and so their wages grow not just
because of seniority.
Labor Force Withdrawals and Wages
Mincer and Polachek (1974) extend the
human capital mode! to account for the pos-
sible depreciation of human capital that may
result from the discontinuity of women's work
experience. They argue that during periods
of labor force withdrawal for child-bearing
and child-rearing, prolonged nonparticipa-
tion in the paid labor market can cause the
skills acquired at school and work to become
less valuable.
The following function adjusts the basic
human capital wage mode} to account for
depreciation or obsolescence effects:
~—~
E`= Es + ~ (rC'—BE`),
i=l
where E`, Es, r, Ci are defined as in Eq.
(1), bi is the depreciation rate of human capi-
tal in year i, and Ei is earnings capacity in
year i.
The total benefits of investments in on-
thejob training increase with the length of
the payoffperiod but decline with the length
of periods of nonparticipation that follow in-
vestments. This suggests that optimal in-
vestment patterns will differ depending on
the continuity of market activities. Contin-
uously employed workers should concen-
trate investments early in their careers.
Workers who interrupt their work careers
will defer investments in on-thejob training
until they reenter the labor market after
completing these activities so as to minimize
the loss from depreciation. Since such work-
ers have a shorter payoff period, their over-
all volume of investment should be lower
than that of workers who remain continu-
ously in the labor force.
Mincer and Ofek (1982) have since re-
vised this initial mode! to account for "res-
toration" or "repair" of depreciated human
capital. They argue that the "reconstruction
of (previously eroded) occupational skills is
more efficient than the construction of new
human capital." That is, it costs less to repair
human capital than to build it. This resto-
ration phenomenon leads Mincer and Ofek
to distinguish the short-run and long-run
consequences of nonparticipation. In the
short run (say, the first year following an in-
terruption), one would expect sharply lower
wages than those received just prior to the
interruption, followed by a period of rapid
wage growth during which human capital is
restored. Thus, the long-run effects on wages
of nonwork time may be considerably smaller
than the short-run effects.
Since the empirical work of Mincer and
Ofek and our own replication of it show that
wage "rebound" following an interruption is
an important phenomenon, it is useful to
consider alternative explanations of it. Cor-
(2) coran (1979) and Corcoran and Duncan (1979)
suggest that time out may lead to a tem-
porary reduction in wages because of tem-
porary mismatches between worker skills and
jobs. Women workers lack complete infor-
mation about job opportunities when they
do return to the labor force, and it takes
time for them to discover jobs that are best
matched to their skills. Employers also have
imperfect information about the productiv-
ity of their new employees, and the learning
process for them is time-consuming.3 One
3 Morgensen (1978), Jovanovic (1979), and Prescott
and Visscher (1980), for example, explain that earnings
rise with experience with a firm because firms learn
about worker productivities in various jobs (instead of
workers' acquiring skills through experience). This
learning process results in the more senior workers
being matched more accurately to jobs commensurate
with their skills than less experienced workers. Better
job matches allow the senior workers to exhibit higher
productivity on average, and if the market rewards
productivity, these differences may account for their
higher average earnings.
OCR for page 174
174
MARY CORCORAN, GREG J. DUNCAN, AND MICHAEL PONZA
common mechanism for this sorting process
is to hire new workers in at low wages but
then to promote them rapidly as they suc-
cessfully complete their probationary pe-
riods. In neither ofthese cases are the work-
ers "restoring" depreciated skills in the
Mincer-Ofek sense. Wage increases accom-
pany improved information of employees
about their job opportunities or improved
information of employers about the produc-
tivity of their employees.
Note that Mincer and Ofek have altered
the original Mincer and Polachek mode!
substantially. Since depreciated human cap-
ital can be restored, it no longer follows that
intermittent workers will necessarily defer
investments in on-thejob training until all
interruptions are over. This decision will de-
pend on the relative sizes of the cleprecia-
tion and restoration effects.4 Similarly, the
relative sizes of these two effects will also
determine the Tong-run wage costs of labor
force withdrawals. If these long-run costs
are small, then depreciation may account for
little of the wage gap between men and
women.
Part-Time Work Experience and Wages
Women are considerably more likely than
men are to work in part-time jobs, a fact that
may lead to considerable differences in the
amount of on-thejob training women ac-
quire and, therefore, in their relative wage
growth. The most general human capital
theories (Heekman, 1976; Blinder and Weiss,
4 A qualification suggested to us by Jacob Mincer is
necessary here: For the intermittent worker, each in-
terruption carries with it a positive probability of not
returning to the labor market. Thus, the expected pay-
off period is diminished by more than the interruption
each time it looms. So although wage loss due to de-
preciation can be made up, the intermittent worker's
decision about whether and when to invest will depend
jointly on the relative sizes of the depreciation and
restoration effects and on the probability of returning
to work.
1976) do not make unambiguous predictions
about the effect of part-time work on human
capital investment and wages, but there are
reasons for believing that less training is ac-
quired in part-time work than in full-time
work. First, because part-time work means
fewer hours in the labor market than full-
time work does, women who expect to work
part time in the future have a shorter ex-
pected work life and hence less incentive to
invest in on-thejob training. In this case
both the overall volume of investment and
the rate of investment would be lower for
those who plan to work part time than for
those who plan to work full time. If current
part-time work patterns are associated with
the likelihood offuture part-time work, then
current part-time workers will be making
fewer investments. Second, if employers
suspect that part-time workers are more likely
to leave than are full-time workers, they
might restrict training opportunities in part-
time work. Employers would be most likely
to restrict opportunities for firm-specific
training. Finally, just as it is argued that
skills depreciate during periods of nonwork,
skills could depreciate more (or appreciate
less) during part-time work than during full-
time work, since part-time work involves
fewer hours of work (i.e., more hours of
nonwork). The depreciation from nonuse
would be greater if the nature of part-time
work precluded workers from maintaining
their market skills. If formal training is
scheduled when part-time workers are not
at work, then there will be less wage growth
resulting from the acquisition of new skills
for them.
Two sources of data with crude direct
measures of on-thejob training do show a
positive relationship between work hours and
training. Duncan and Hoffman (1979) found
with the 1976 wave of the PSID that adult
workers aged 18 to 64 who worked less than
20 hours per week reported training periods
attached to their jobs that were only about
half as long as those of workers working be-
tween 40 and 50 hours. Stafford and Duncan
OCR for page 175
WORK EXPERIENCE, [OB SEGREGATION, AND WAGES
175
(1979) found qualitatively similar, although
less statistically significant, differences in
training by labor supply for workers in the
1975-1976 Time Use Study.
Labor Force Withdrawals and Job
Segregation: Human Capital Explanations
The 1974 Economic Report of the Presi-
dent speculated that sex differences in pat-
terns of work participation may be the cause
of the sex segregation of jobs. This line of
reasoning has been extensively developed
by ZelIner (1975) and Polachek (1976, 1979,
19811.5 Since Polachek's explanation sub-
sumes Zeliner's model, we will concentrate
on his mode! in the following discussion.
Polachek (1981) defines atrophy as the loss
of earnings potential that occurs when skills
are not continuously used. He shows that if
the cost of labor force withdrawals (the atro-
phy rate) varies across occupations, and if
lifetime labor force participation differs across
individuals, then a worker will choose "that
occupation which imposes the smallest pen-
alty, given his desired lifetime participa-
tion." This model treats the "lifetime as a
unit."6 Thus, this mode! implicitly assumes
that workers tend to work in the same sort
of occupation throughout their lives or at
least over long periods of time. Lifetime work
participation is assumed to be exogenously
determined, and atrophy rates are assumed
to vary across occupations.
This mode! provides a human capital ex-
planation for the sex segregation of the labor
market. If work skills do atrophy during
s England (1981, 1982) provides an extensive discus-
sion of these models. This section is informed by her
work.
6 Polachek notes that "this assumption can be relaxed
by posing the problem within a dynamic control frame-
work" but goes on to say that "even within such a
framework the same conclusions hold for occupations
chosen at a given stage of the life cycle" (Polachek,
1981, p. 64~. Note that this relaxation still implies oc-
cupational immobility over a life-cycle stage.
withdrawals from the labor force, then it is
rational for women who expect to take time
out from the labor force to work in fields
where there is less chance of atrophy i. e.,
in fields with low depreciation rates but also
with low returns to experience. Thus, such
women wid experience less atrophy than will
women who expect more continuous work
participation. By selecting jobs that are easy
to leave and reenter, women can thus more
easily combine the clual demands of career
and family. Polachek's mode} can explain sex
segregation only if typically "female" jobs
are those where there is the least atrophy.
Note that if depreciates! skills can be re-
stored (as Mincer and Ofek argue), this
weakens the force of Polachek's arguments.
Webster's defines atrophy as "a wasting
away or progressive decline." Thus, the cas-
ual reader might assume that Polachek's
atrophy rate is equivalent to Mincer and
Polachek's depreciation rate. But atrophy,
as defined by Polachek, picks up two things—
depreciation (i.e., reduction in work skills
due to nonuse) and forgone appreciation (i.e.,
the loss in expected earnings growth due to
missing a year of work).7
Depreciation and the growth of earnings
with experience are quite different proc-
esses. Depreciation implies that the level of
work skills is lower following an interruption
7 We are grateful to Siv GustaEson for first pointing
out this distinction to us in a personal conversation in
1978. England (1981, 1982) is the first author who clearly
makes this distinction in a published paper. England
first referred to the loss in expected earnings growth
as "forgone appreciation," and provides an excellent
discussion of Polachek's models.
Zellner's explanation of the sex segregation of oc-
cupations rests solely on forgone appreciation. England
(1981) points out: "Zellner assumes that occupations
can be divided into those that offer high initial salaries
and flat earnings profiles and into those with low initial
salaries and steep earnings profiles. Women, because
of their shorter expected work lives, will be more likely
to maximize lifetime earnings in the occupations with
high initial salaries and flat wage growth i.e., in 'fe-
male' occupations."
OCR for page 176
176
MARY CORCORAN, GREG [. DUNCAN, AND MICHAEL PONZA
than it was just prior to that interruption. If
this earnings loss is long lasting, it is obvious
why women who expect prolonged labor force
withdrawals should enter fields with low de-
preciation rates. Polachek's mocle! also im-
plies that women who expect prolonged
withdrawals should enter fields with high
initial salaries but fairly flat earnings growth
rates. This second decision only makes sense
if we assume that, all else equal, jobs with
high earnings growth pay less initially than
jobs without such earnings growth. (Eng-
land [1981] makes this point quite clearly.)
To summarize, Polachek provides an in-
genious human capital explanation for job
segregation. If Polachek's general mode} is
correct, then women who anticipate pro-
longed nonwork time shouicl work in fields
with low atrophy rates and should experi-
ence less depreciation and less wage growth
than do otherwise similar workers. Pola-
chek's model also has several implications
for the nature of typically female and male
occupations. First, since women's choice of
a "female" or"male" occupation reflects life-
time participation plans, we would expect
that the sex typing of women's occupations
change little over a prolonged period of time.
In addition, we should find that deprecia-
tion and/or earnings growth will be lower in
"female" occupations and that women who
expect discontinuous careers will choose
"female" rather than "male" occupations be-
cause discontinuity is penalized less. Be-
cause of this choice, women with discontin-
uous work careers will be concentrated in
"female" occupations. Occupational immo-
bility and low wage growth are also pre-
dicted by any job segregation model that
presumes that women are locked into a set
of female-dominated jobs that do not pro-
vide productivity-enhancing experience.
WOMEN'S WORK AND OCCUPATIONAL
HISTORIES
The human capital models summarized
above predict that women's low wages result
from a low overall volume of work, inter-
mittent work participation, and part-time
work. Most job segregation models, human
capital or otherwise, implicitly assume con-
siderable immobility between "female" and
"male" jobs. As a first step toward testing
these models, we use 13 years of data from
the PSID to assess the accuracy of these
descriptions of women's work behavior and
occupational immobility.
Patterns of Labor Supply, 1967 to 1979
Every year PSID respondents report their
own and their spouse's work hours. Cor-
coran et al. (in press) examined labor supply
for adult men and women, aged 23 to 47 in
the first year of the panel, who lived in their
own households.8 As expected, there were
dramatic differences between the sexes in
the frequency and regularity of work and in
the extent of part-time or part-year work.
Differences between the races were much
less dramatic within the groups of men and
women. Black women acquired more ex-
perience than white women did, while black
men acquired less of it than did white men.
Between 70 and 80 percent of the two
groups of women were absentirom the labor
force for at least 1 of the 13 years.9 The
~ In terms of the PSID sample, this group consists
of all individuals who were household heads or wives
in each of the 13 years. Eliminated from this analysis
are children and the small group of other relatives of
the household head (e. g., brother or sister). The lower
age restriction was imposed to avoid sample selection
problems associated with the decision to leave the pa-
rental home and form one's own household. The upper
age restriction eliminated from the sample individuals
who would have reached the early retirement age of
62 by the end of the panel period. The results are
reported in Corcoran et al. (in press).
9 We used 250 hours during a calendar year to define
whether an individual was in or out of the labor force
during that year, and we use 1,500 hours to separate
part- from full-time workers. This procedure has its
disadvantages. An individual with a 40-hour-per-week
job who drops out of the labor force altogether for six
months out of a calendar year will be classified as a
part-time worker during that year without having a
spell of nonwork. In one sense this individual was a
OCR for page 177
J
WORK EXPERIENCE, IOB SEGREGATION, AND WAGES
177
comparable fractions for white and black men
were about 10 and 16 percent, respectively.
Even when they did work, women were
much less likely to work full time. Less than
one-tenth of these adult women worked full-
time during the entire 13-year period.
The total volume of work experience ac-
quired by men was much higher than for
women. On average, men worked in almost
twice as many of the years as women did,
in nearly twice as many weeks as women
did, and for nearly three times as many hours
as women did during this period.
The part-time nature of the work of women
was highlighted when we examined hours
worked per week and weeks worked per
year during the work spells. The average
work week of men exceeded 40 hours,
amounting to 46 hours for white men and
43 hours for black men. In contrast, white
and black women averaged 36 hours per week
during work spells. Similar differences
showed up in the number of weeks worked
per year, with men averaging 47 weeks and
women averaging 42 weeks. As a result, the
tote] number of hours averaged by men dur-
ing their work spells was almost twice as
high as for women during their spells.
Patterns of Occupational Segregation,
1975 to 1979
Both Polachek's (1981) model and seg-
mented labor market models assume little
mobility between "male" and "female" oc-
cupations over a prolonged period of time.
Yet England (1982) reports that the corre-
lation between percent female in detailed
census occupation coding of first job and
percent female in detailed coding of 1967
job is only .39 for women aged 30 to 44 years
in 1967. This suggests there may be consid-
erable mobility between "male?' and "fe-
male" job sectors.
We further tested this assumption of in-
tersectoral immobility by calculating pat-
terns of occupational segregation over the
years 1975 to 1979. Our measure of occu-
pational segregation is based on 2-digit oc-
cupation and 2-digit industry categories.
Thus, it has the advantage of accounting for
both occupational and industrial segmen-
tation by sex.~° We define a female-domi-
nated job as an industry-occupation group
with more than 50 percent women workers.
Industry-occupation groups with less than
50 percent female workers are designated
male-dominated jobs.
To investigate the dynamics of job seg-
regation over the period from 1975 to 1979,
we selected a sample of women aged 23 to
57 in 1975 who worked in the first and last
years of that period and who may or may
not have worked during the three years in
between. ii About 70 percent of white women
workers held female-dominated jobs in 1975.
If job segregation was completely rigid, then
we would expect to observe that same frac-
tion spending all of their working years in
jobs dominated by women. Table 10-1 shows
that this is clearly not the case. Only half of
the white women spent all of their working
years in the five years between 1975 and
1979 in jobs dominated by women. Less than
one-sixth of these white women spent all of
their working years in jobs dominated by
men, leaving more than one-third who
switched job types at least once. Switches
between female- and male-dominated jobs
for black women were almost as common
full-time worker and in another sense this individual
experienced a spell of nonwork during that year. Our A Our procedure for determining whether a given
measure considered part-time workers to be those either job was "female-dominated" or "male-dominated" is
working a limited number of hours per week or those detailed in Corcoran et al. (in press).
working during only part of the year. Our measure of ii As before, the sample consists of household heads
nonwork spells required that such spells be long enough ' -' ' ~' ' ' ''
to take an individual away from work for virtually an
entire year.
and wives in this age range and thus excludes a small
number of adults who are related to the head of the
household in some other way.
OCR for page 178
178
MARY CORCORAN, GREG I. DUNCAN, AND MICHAEL PONZA
TABLE 10-1 Dynamics of Occupational Segregation
Subgroup
Women Who Worked at Least 2S0
Hours in 1975 and 1979
Occupational Change
Fraction spending all working years in
female-dominated jobsa
Fraction spending all working years in
male-dominated jobs
Men
All White
White Black
Black
.07 .18
(1,563) (606)
All
.08
(2,169)
Fraction switching at least once in
either direction
Fraction of those in female-dominated
jobs initially who switched to male-
dominated jobs
.51
(871)
.15
(871)
34
.
(871)
.31
(647)
.61
(538)
.09
(538)
.31
(538)
.25
(423)
.52
(1,409)
.14
(1,409)
34
.
(1,409)
.30
(1,070)
.79
(1,563)
.15
(1,563)
.60
(207)
.67
(606)
.18
(606)
.37
(106)
.78
(2,169)
.16
(2,169)
.56
(313)
Fraction of those in male-dominated
jobs initially who switched to female- .44 .55 .45 .09 .13 .09
dominated jobs (224) (115) (339) (1,300) (456) (1,756)
NOTES: Table reads: 51 percent of the 871 white women who worked at least 250 hours in 1975 and 1979 spent
all of their working time in female-dominated jobs.
The number of observations is given in parentheses below each estimate.
a A job is designated as female dominated if the percentage of women comprising it is greater than or equal to
50. Otherwise it is designated as male dominated.
SOURCE: Panel Study of Income Dynamics.
31 percent of the black women were codes!
as switching from one type to the other.
These figures on the extent of switching
between job types mix together both kinds
of changes. A more relevant statistic on the
issue of whether women who take female-
dominated jobs remain in them is the frac-
tion of women who began in female-clomi-
natec! jobs and switched out of them. That
fraction is 31 percent for white women and
25 percent for black women.
These figures on switches between job
types deserve careful scrutiny. On the one
hand, they are likely to understate the true
amount of movement between job types be-
cause the time span over which such changes
can be observed is limited to only five years. i2
i2 Women who worked continuously in the same sec-
tor between 1975 and 1979 but switched in 1980 or
had switched in 1974 are classified here as persistent
residents in one sector. Also, while most of the women
in this sample worked in every one of the five years,
In addition, our classification procedure for
identifying female- and male-dominated oc-
cupations is a crude one and undoubtedly
misses some true switches that would be
caught with a more refined set of occupa-
tional and industrial codes. On the other
hand, errors in the coding of occupation and
industry may create the appearance of a
switch when in fact there was nones It is
impossible to say whether the net effect of
these considerations is to increase or to de-
crease the estimated extent of switching be-
some did not work in some of the middle three years,
giving them fewer than five years in which a switch
can be observed.
i3 Appendix Table B.l in Corcoran et al. (in press)
sheds some light on this by showing comparable mo-
bility figures for the case when male-dominated jobs
are defined as less than 40 percent female, and female-
dominated jobs are defined as greater than 60 percent
female. Hess mobility is found with this more restric-
tive definition, but the extent of mobility is still sub-
stantial.
OCR for page 179
WORK EXPERIENCE
, . _
SEGREGaTION. AND WAGES
179
tween job sectors. However, it is almost cer-
tainly true that the extent of switching is
substantial, a fact that is inconsistent with
Polachek's assumption of occupational im-
mobility and with any other labor market
mode} based on rigid segmentation by sex.
This result also suggests that analysts should
be wary of using a woman's current occu-
pation as a measure of her past occupational
history.
WORK HISTORY AND WAGES:
EMPIRICAL EVIDENCE
The Depreciation Effect
Empirical evidence on whether deprecia-
tion lowers wages consists mostly of cross-
sectional data where earnings of different
individuals with different work histories are
compared, after statistical adjustment to make
the individuals as similar as possible. This
evidence has produced contradictory results
about the size of the depreciation effect-
i.e., about the extent to which wages de-
cline with time out of the labor force once
one controls experience and tenure. Mincer
and Polachek (1974) reported Mat 1967 wages
dropped by 1.2 percent per year out of work
for white married women aged 30 to 44 with
children. Sandell and Shapiro (1978) repli-
cated the Mincer-Polachek analysis after
correcting for coding errors in women's re-
ports of employment behavior. They re-
ported that wages cleclined only 0.4 percent
per year of nonparticipation and that this
effect was insignificant. Corcoran (1979) rep-
licated the Mincer-Polachek analysis for a
national sample of wives aged 30 to 44 with
children, taken from the 1975 PSID and ob-
tained similar results to those of Mincer ancl
Polachek. But Corcoran (1979) and Cor-
coran and Duncan (1979) also reported that
the decline in wages was much smaller (0.6
percent per year out of work) for working
women in a broader age range (18 to 64
years). These results suggest that wages of
married women aged 30 to 44 are more af-
fected by labor force withdrawals than are
wages of women in a broader age range.
A recent paper by Mincer and Ofek (1982)
suggests that some of these inconsistencies
in past research arise because cross-sec-
tional analyses tend to confound the short-
run and long-run effects of nonparticipation.
Mincer and Ofek use eight years of National
Longitudinal Survey (NLS) data to explore
how time out of the labor force affected wage
growth for a sample of women aged 30 to
44 in 1967 who were married sometime dur-
ing 1967 to 1974. They found a large short-
run loss in wages immediately following an
interruption (ranging from 3.6 to 8.9 per-
cent per year out of work), followed by a
period of rapid wage growth. Estimates of
long-run wage losses were moderae (0.4 to
1.1 percent per year out of work). A repli-
cation and extension ofthis analysis was con-
ducted by Corcoran et al. (in press) using
13 years of information from the PSID. Short-
run depreciation effects were estimated to
range from 2.5 to 4.7 percent per year de-
pending on the exact form of the mode! and
on the definition of the sample when the
age range was identical to the one used by
Mincer and Ofek (30 to 44~? and the effects
were estimated to be similar in magnitude
when the age range was extended to be-
tween 23 and 47 years. Long-run deprecia-
tion was estimated to be between 1.0 and
1.5 percent per year in the replication, with
some of these coefficients not statistically
significant at conventional levels.
Ibis recent work reconciles the disparate
estimates of depreciation. Ike past analyses
of wage effects of withdrawals based on cross-
sectional data are likely to pick up both short-
run and long-run effects. Married working
women aged 30 to 44 who have interrupted
work are likely to have recently returned to
the labor force, and so short-run effects may
have a large weight in analyses run on this
group (e.g., that of Mincer and Polachek,
19741. Analyses run on women in a broader
age range (e.g., that of Corcoran, 1979; and
that of Corcoran and Duncan, 1979) are likely
to put more weight on the long-run effect
OCR for page 180
180
MARY CORCORAN, GREG r. DUNCAN, AND MICHAEL PONZA
-
and so will provide lower estimates of de-
preciation.
The Restoration Effect
In their original paper, Mincer and Po-
lachek (1974) suggested that the optimal
timing of investment in on-thejob training
would differ depending on the continuity of
market activities. In particular, workers who
interrupt their work careers for nonmarket
activities will defer investments in on-the-
job training until they reenter the labor mar-
ket after completing these activities so as to
minimize the loss from depreciation. Min-
cer and Ofek (1982) have considerably re-
vised this hypothesis by arguing that "de-
preciated" or "erodecl" human capital can
be cheaply and rapidly "restored" soon after
labor market entry. They show that postin-
terruptior~ wages grow at roughly 2.5 per-
cent per year of experience, on average, and
that growth rates in the first year following
an interruption range from 5.8 to 6.4 per-
cent per year depending on the exact spec-
ification of their model. This growth rapidly
erases estimated short-term losses from the
depreciation associated with short spells out
of the labor force and is much larger than
the wage growth of comparable continuous
workers. They further demonstrate that
growth in tenure accounts for less than half
of this wage growth and interpret this to
mean that the remainder is due to growth
(repair) of general training.
The nature and causes of wage rebound
following work interruptions are crucial ele-
ments in understancling the wage conse-
quences and job choice of female labor sup-
ply patterns. If depreciation is quickly
repaired, it no longer follows that intermit-
tent workers will defer investments in on-
thejob training until all interruptions are
completed. Thus, the observation that the
wages of women grow more slowly in the
years following the completion of schooling
because of the reduced incentives to invest
in human capital may no longer hold. It also
weakens the plausibility of the reasoning that
intermittent workers will concentrate in fe-
male jobs.
The Mincer and Ofek (1982) estimates of
restoration came from the estimation of a
cross-sectional wage equation. Corcoran et
al. (in press) were able to use more complete
information about the amount and sex typ-
ing of work experience before and after work
interruptions and also estimated a wage-
change equation a specification with sev-
eral statistical properties that make it pre-
ferred to a wage-ZeveZ equation. Since this
work has only recently been completed and
addresses many of the important issues con-
sidered in this paper, we summarize our
analysis here. Readers interested in the de-
taiTs are referred to Corcoran et al. (in press).
We selected the adult women in the PSID
sample and used the 13 years of work history
obtained for them. We developed our wage
equation by identifying the first (F) and last
(L) wage observation for all women who
worked at least 2 of the 13 years. 14 A cross-
sectional wage equation at time F would be
of the form:
ZnW`F = aOF + alFSiF + a2FeOiF
+ 0~3FhOiF + ~FXiF + IF + IF ~ (3)
where W`F is the wage rate of the ith indi-
vidual in time F. aOF is a constant in time
F. SiF is the education level of the ith indi-
vidual in time F. eOiF is years of work ex-
perience for the ith individual in time F. hoiF
is years of nonwork for the ith individual in
time F. XiF is a vector of observed produc-
tivity-related characteristics for the ith in-
dividual in time F. ZiF iS a vector of unob-
servable individual-specific productivity-
related characteristics, workplace charac-
teristics, and labor market differences for
the ith individual in time F. and ifs iS the
14 Throughout this section we use 250 hours as the
cuto~point to distinguish individuals in the labor force
from those out of the labor force. The sensitivity of
these results to change in this definition are detailed
in Corcoran et al. (in press).
OCR for page 181
WORK EXPERIENCE, JOB SEGREGATION, AND WAGES
181
stochastic disturbance term for the ith in-
dividual in time F.
PSID information on the work history be-
tween F and L can be used to distinguish:
en, years of work experience accumulated
between F ant! L since the last completed
interruption; e*, years of work experience
accumulated between F and L prior to the
most recent completed interruption; hi' years
of nonwork between F and L during most
recent completed interruption; and h*, years
of nonwork accumulated between F and L
prior to most recent completed interrup-
tion.
The cross-sectional wage relationship at
time L (allowing the parameters to change)
is given by
InWiL, = aloe + Apsis +
(X2reoiF + a3~hOiF +
a4e*i + (X5eLi +
a6h*i + Aphid +
~UrXi~ + ~~ZiL. + Din (4)
The short-run depreciation and rebounc!
effects are given by parameters a7 and as,
respectively. Subtracting Eq. (3) from Eq.
(4), suppressing the subscript i, denoting
changes from F to L as "A," and adding and
subtracting Arose, OAF, ant! U~ZF results
in the following general equation for wage
change:
AlnW= Mao + a~r/`s +
AalSF + ~a2eOF +
~a3hOF + a4e* + Uses +
a6h* + a7hL, + ,u~,AX +
/\ I1XF + BUZZ + A~TrZF
+~.
(5)
If one assumes that the cross-sectional ef-
fects of the explanatory variables are invar-
iant between F ant! L and, further, that the
unmeasured characteristics remain constant
for the same individual, then the wage change
equation simplifies to:
AInW= auras + a4e* + aSeL,
+ ash* + a7hl + ,u~LliX + AD (6)
Although the dependent variable in Eq. (6)
is wage change rather than wage level, the
parameters on the experience variables (a4
- a7) correspond to the parameters in the
cross-sectional Eq. (4~. The key advantage
to the change formulation is that estimates
of these parameters are free from the sta-
tistical problems caused by retrospective re-
ports and by unchanging, unmeasured var-
iables correlated with the included
(measured) explanatory variables. An addi-
tional advantage is that many more women
meet the requirements of working at least
2 of the 13 years than work in a single year,
and, therefore, selection bias problems are
much less severe in estimating change Eq.
(6) than in estimating a cross-sectional equa-
tion. 15
Table 10-2, columns 1 and 2, shows es-
timates of wage-change Eq. (6), which is the
longitudinal analogue to the Mincer-Ofek
cross-sectional equation. The work segment
following the most recent interruption cent
was entered quadratically to allow for a more
rapid growth at first. Results show the es-
timated rate of wage growth immediately
following the last interruption to be a little
over 5 percent per year for white women
and 8 percent for black women, with the
rate of growth declining to zero after about
10 years for both groups, which is close to
the maximum observer! value for the eL var-
iable in the sample. Depreciation during the
most recent interruption is estimated to be-
tween 4 and 5 percent per year, so the initial
wage rebound following an interruption more
than makes up for the wages lost during a
year out of the labor force.
Effects of Prospective Interruptions on
Wage Growth
Since the profitability of investments is
affected by the length of time over which
is See Corcoran et al. (in press) for more detailed
discussions of the development of the wage-change
equation and of selection bias adjustments.
OCR for page 182
182
MARY CORCORAN, GREG I. DUNCAN, AND MICHAEL PONZA
TABLE 10-2 Basic Wage Growth Regression
Independent Variable White
h*: years out of labor force prior to most recent
interruption
ha: years out of labor force during most recent
interruption
e*: years in labor force prior to most recent reentry
Black
016
`.030y
046**
t.016'
.030**
`.008'
.080**
`.026y
.0041*
`.0019y
White
Black
-.015
`.~o'
-.070**
`.020y
.023*
`.011'
.077*
<.032'
- .0045*
(-arm)
-.124
`.103y
.osst
`.032'
- .006
`.008y
.ooos
.0090)
.035
521
en: years in labor force during most recent spell
e 2
L
NT79: Did not work in 1979
NT79*h,
NT79*el.
NT79*e~,2
R2 (adjusted)
Number of observations
.016
`.029)
038**
`.013'
.012t
(-~
.0s2*
`.o~'
.0027t
.0016)
.021 .057
837 521
.013
.029
035*
.015
.012
(.oo9)
.05
.027
0025
(.0015
017
(.o99)
030
.029
.043
(.056
008
.0056)
.024
837
NOTE: Standard errors are in parentheses.
* Significant at .05 level.
** Significant at .01 level.
Significant at .10 level.
SOURCE: Panel Study of Income Dynamics.
benefits are received, the human capital
mode} predicts that otherwise identical
workers will invest less if they anticipate
labor market withdrawals than if they do
not. This assumption is also implicit in Po-
lachek's argument that workers who antic-
ipate time out will choose occupations with
low atrophy rates. Sandell and Shapiro (1980)
test this proposition by estimating whether
NLS women who expected to be out of the
labor force at age 35 had flatter experience-
earnings profiles before then than did women
who expected to be working at age 35. Al-
though most of their key parameter esti-
mates are in the expected direction, none
are significant at the 5 percent level for white
or black women.
This hypothesis is an important one for
the human capital model; it deserves testing
in the context of the wage-change models
developed here. Since we know which
women in the PSID sample were not work-
ing at the end of the 13-year period, we can
test directly whether such workers'jobs pro-
vided them lower wage growth and lower
depreciation. In contrast to the self-re-
ported intentions of respondents used in the
articles listed above, this procedure tests for
the effects of actual labor force behavior in
period t + 1 on wage profiles in period t.
We did this by creating a dummy variable
(NT79) for whether did not work in 1979,
interacting this dummy with he, en, and
e2, and adding these four variables to the
basic wage-change Eq. (61. The results of
this analysis are reported in Table 10-3, col-
umns 3 and 4. In general, white women who
did not work in 1979 had the same wage
increment for additional years of experience
as women who did work in 1979 and similar
wage loss with time out as did otherwise
similar white women who were working in
1979. But one result for black women does
conform with human capital predictions.
OCR for page 183
interruption
h,: years out of labor force
during most recent
interruption
Years of full-time e*
Years of part-time e*
Years of full-time e'
(Years of full-time eJ-squared
Years of part-time en
.034
(.013)
.016*
(.008)
.001
(.011)
.029
(.008)
.001
(.010)
.037*
(.015)
.024*
(.008)
.018
(.011)
.030
(.010)
.017
(.013)
-.035**
(.013)
.007
(.007)
- .002
(.011)
.062**
(.021)
- .0036*
(.0018)
- .021
(.026)
(Years of part-time e-squared .0016
(.0025)
Years of full-time en in male-
dominated jobs
Years of part-time e, in male-
dominated jobs
Years of full-time en in female-
dominated jobs
WORK EXPERIENCE, TOB SEGREGATION, AND WAGES
183
TABLE 10-3 Effects of Part-Time Work and Female-Dominated Work on Wage Growth
Independent Variable White Black White Black White
h*: years out of labor force .023 - .005 .022 - .026 .021
prior to most recent (.029) (.030) (.029) (.029) (.029)
-.018
(.029)
.049
(.015)
.034
(.008)
.023*
(.011)
.037
(.024)
.008
(.0021
.150*
(.029)
.014
(.003)
.036
(.013)
.005
(.008)
.003
(.012)
.026*
(.012)
.048*
(.015)
.015
(.001)
.008
(.011)
.008
(.016)
Years of part-time eL in female- - .004 - .003
dominated jobs - (.012) (.014)
Number of cases 837 521 837 521 837 521
R2 (adjusted) .025 .049 .028 .096 .023 .044
- .009
(.019)
.020*
(.009)
.024
(.028)
.021*
(.011)
* Significant at .05 level.
** Significant at .01 level.
SOURCE: Panel Study of Income Dynamics.
Black women who were not working in 1979
exhibited no wage loss during prior labor
force withdrawals. i6 Their past wage return
to experience, however, did not differ from
i6 The depreciation estimate for black women who
did not work in 1979 is the sum of the coefficient on
he term (-.070) and the coefficient on the NT79*h~
term (+.059).
that of otherwise similar black women who
were working in 1979.
Part-Time Work Experience and
Intermittency
Corcoran (1979) and Corcoran and Dun-
can (1979) included a retrospective measure
of experience in a cross-sectional equation
for a national sample of women. They report
OCR for page 184
- 84
MARY CORCORAN, GREG 1. DUNCAN, AND MICHAEL PONZA
that part-time work experience was signifi-
cantly less valuable than full-time work ex-
perience.
Jones and Long (1979) used data from the
NLS to estimate two cross-sectional wage
equations that included interactions be-
tween work experience segments and
whether the work was for part of a week.
Although the signs of the coefficients they
estimated were consistent with the hypoth-
esis that part-time work leads to slower wage
growth, only two of the 12 coefficients were
statistically significant at conventional lev-
els. Their measures of the part-time nature
ofthe work segments were very rough, how-
ever, and may have biased some of the coef-
ficient estimates.
Corcoran et al. (in press) also investigated
the effects of part- and full-time work on the
wage growth of women workers with some
simple modifications to their basic wage-
change equation. In Eq. (6), full- and part-
time years in the e* and en segments were
not distinguished. Since the volume of work
hours was ascertained for each of the 13 years
under investigation, that information can be
used to classify years of experience that in-
volved part-time work Jess than 1,500 hours)
and full-time work (1,500 hours or more).
Four variables were formed with this infor-
mation: (1) the number of years of e* that
were part-time (e*-part), (2) the number of
years of e* that were full-time (e*-ful0, and
(3) the number of years of en that were part-
time (er-part), and (4) the number of vears
of en that were full-time (er-ful0. This de-
composition of e* and e~yields the following
equation:
A1,nW= a,L,/\S +
a6h* + a7hL, +
a~e*-part +
age*-full +
a~OeL,-part +
a~eL-full +
FLAX + AD. (7)
As with the more basic measures of e* and
en, all four of these new variables are ob-
tained in each of the annual interviews and
do not rely on retrospective reports by either
women workers or their husbands.
The results of the estimation of the aug-
mented wage-growth equation are shown in
Table 10-2, columns 1 to 4. Full-time work
floes indeed appear to be associated with
significant wage growth, while part-time work
does not. When the two measures of en are
entered linearly, the wage growth associ-
ated with full-time experience is positive and
significant for both white and black women,
while the wage growth associated with years
of part-time work in the most recent spell
of employment was insignificant for both
groups of workers. With years of experience
prior to the most recent spell of nonwork
(e*), the full-time work variables have larger
coefficients than do the part-time variables,
although these differences were not signif-
icant at conventional levels. A parabolic
specification for the en measure gives ex-
pected results for white women there is a
parabolic rebound for full-time but not part-
time work. For black women, there is par-
abolic wage rebound for part-time work-
a puzzling result.
Sex Differences in Work History and the
Sex-Based Wage Gap
Two sets of analysts have extensively ex-
amined the relationship between work his-
tory and the sex-based wage gap on a na-
tionally representative sample of women.
Mincer and Polachek (1974) estimated that
sex differences in experience and time not
working accounted for about 45 percent of
the wage gap between employed married
men and women aged 30 to 44 years in 1966.
About half of this difference was due to the
depreciation effect. i7 Corcoran and Duncan
}7 Mincer and Polachek used data from two different
sources to make this comparison. This led to some
inconsistencies between male and female variables. The
sample of men was taken from the 1966 Survey of Eco-
nomic Opportunity (SEO). This survey does not meas-
ure either work experience or tenure directly. Men
OCR for page 185
WORK EXPERIENCE [OB SEGREGATION AND WAGES
. .
(1979), using a broader age range and a more
extensive list of work history measures, found
that sex differences in work history ac-
counted for between one-third and two-fif;chs
of the wage gap between working white men
and working women aged 18 to 64 in 1975.
This occurred largely because women had
acquires] less tenure and were more likely
to have worked part time. The depreciation
effect and intermittence clid little to explain
the wage gap between women and white
men.
Work History and Occupational
Segregation: Empirical Evidence
Polachek's (1981) argument implies that
women choose "female" jobs because such
jobs penalize discontinuous labor force par-
ticipation less than "male" jobs do. This ex-
planation presumes that there is immobility
between occupations; that depreciation and/
or wage growth are lower in "female" jobs
than in "male" jobs and that women who
expect discontinuous careers are concentra-
ted in "female" jobs. We have already demon-
strated that there is substantial mobility be-
tween "male" and "female" job sectors over
a five-year period a result that contradicts
a basic assumption of Polachek's explanation.
The empirical evidence presented by Po-
lachek for this argument is indirect. Pola-
chek (1981) has shown that the probability
of currently working in a given occupation
(keened by 1-digit census categories) is af-
fected by years out of the labor force and
that the size of this effect differs by occu-
pation. He has also demonstrated that the
relationship between wage growth and years
out of the labor force (home time) differs
across occupations (i.e., occupations have
different atrophy rates). He has further shown
were assumed to have no interruptions. These two re-
sults are not inconsistent. In both cases, sex differences
in work history explain a large but not major part of
the sex-based wage gap. Sample differences likely ac-
count for differences in the importance of depreciation.
185
that there is a negative correlation between
the effect of years out of the labor force on
the probability of working in an occupation
and the atrophy rate in that occupation.
Even if we ignore the issue of mobility
across occupations, Polachek's evidence does
not unambiguously support his hypothesis.
Take Polachek's first finding that home time
affects the probabilities of currently being
in a particular occupation. This finding can
only explain sex segregation of jobs if women
with extensive home time were more likely
to work in female-dominated occupations than
were otherwise similar women without ex-
tensive home time. Polachek used his es-
timates of these effects to obtain a projected
population-wicle occupational distribution
for women 30 to 44 if they had worked con-
tinuously since school completion. He re-
ported that the proportion of women profes-
sionals anal managers (currently male-
dominated fields) would increase and that
the proportion of women in household and
service work (currently female-dominated
fields) would decrease. On the other hand,
his figures indicated an increase in the pro-
portion of women employed in clerical work
(a female-dominated field), a decrease in
women employed in crafts (a male-domi-
nated field), and a decrease in the propor-
tion of women in sales (an integrated field). i8
i8 Probably the best way to evaluate the quantitative
importance of this evidence is to estimate the extent
to which occupational sex segregation would be re-
duced if women's home time were zero i.e., if men
and women had the same work participation patterns.
We estimated this by applying Duncan and Duncan's
(1955) segregation index to Polachek's sample. This in-
dex measures "the minimum proportion of one group
that would have to be shifted for its occupational dis-
tribution to be equal to that of the other." The seg-
regation index for Polachek's sample is .50. Then we
calculated this segregation index on the projected oc-
cupational distribution calculated by Polachek under
the assumption that women worked continuously. If
his theory is correct, occupational sex segregation should
be considerably reduced. Under the assumption that
women do not withdraw from the labor force, the seg-
regation index is .48 a reduction of only 2 percent.
OCR for page 186
186
MARY CORCORAN, GREG I. DUNCAN, AND MICHAEL PONZA
England (1982) investigated the effect of
home time on sex typing of current occu-
pation more directly. She reports that the
sex composition of most recent occupation
and the sex composition of first occupation
were uncorrelated with the proportion of
total time employed since school completion
for white women aged 36 to 50 years in 1973. i9
This result does not suggest a strong link
between labor force discontinuity and the
sex typing of current or first job.
Now we turn to Polachek's second find-
ing that the detrimental wage effects of
home time vary across occupations. He cal-
culated this by regressing the difference be-
tween 1972 and 1967 wages on home time
and other variables expected to affect wages.
Unlike most economic studies of wage dif-
ferentials based on the human capital model,
Polachek examines dollar changes in wages
rather than percentage changes. If all oc-
cupations had identical percentage wage de-
creases per year out of the labor force, Po-
lachek would likely obtain differences in
dollar wage change, with highly paid oc-
cupations showing greater decline. Indeed,
in Polachek's analysis, dollar wage changes
are most negative for professionals, crafts-
people, and managers the three highest-
paid occupations.
In order to study wage change between
1967 and 1972, Polachek must restrict anal-
ysis to women who reported a wage in 1967
and in 1972. Thus, Polachek's sample is cho-
sen on the basis of work behavior. This could
possibly lead to selection bias problems when
estimating effects of work behavior on wage
change.20
A further problem is that Polachek's es-
timate of effects of home time will be dom-
inated by short-run effects, since he restricts
analysis to a five-year period and only ex-
amines the effects of withdrawals during that
period. If lost skills were rapidly restored
(as Mincer's and Ofek's t1982] and our re-
sults suggest), then Polachek's estimates will
considerably exaggerate the lifetime costs of
time out. Finally, note that Polachek's atro-
phy estimates pick up both depreciation and
the forgone appreciation effects of fewer years
of experience.
Even if Polachek's evidence that occu-
pations differ in atrophy rates were correct,
this would only explain sex segregation if
there were less depreciation of skills and
lower returns to experience in femaTe-dom-
inated occupations. England (1981, 1982)
tested this assumption using both the NLS
sample of mature women and a sample of
women in a wider age range from the PSID.
England's analyses have the two advantages
of Polachek's analysis. She looks directly at
the relationship between home time and sex
typing of current job, and she examines de-
preciation and returns to experience sepa-
rately. England regressed the natural log-
arithm of wages on experience, education,
time out, and percent female in current oc-
cupation (coded at 3-digit census level), and
tested for significant interactions between
experience and percent female and between
time out and percent female. She reports
that neither the depreciation rate nor re-
turns to experience were affected by per-
cent female in current occupation. This is
fairly strong evidence against Polachek's ex-
planation.
Both Polachek's and England's empirical
tests of the Polachek argument presume
considerable immobility between occupa-
~9 England (1981) reports that Wolfe and Rosenfeld
(1978) present some evidence that suggests a weak link
between home time and sex composition of occupation.
20 This sample selection bias is a general problem for
analysts of women's wages. At any point in time only
about half of all adult women are in the labor force.
Restricting the sample to women who worked in two
specific years as Polachek does eliminates even more
women from the sample. The wage-growth analysis of
Corcoran et al. eliminates only about one-fifth of the
sample, since it requires only that women work at least
2 of the 13 years under investigation. When we rep-
licated our wage-change model for women who worked
in 1975 and 1979, the results were inconsistent with
results from the larger and less restricted sample.
OCR for page 187
WORK EXPERIENCE, [OB SEGREGATION, AND WAGES
187
tions over time. Whether their use of cur-
rent occupation as a measure of occupational
history is appropriate depends upon the va-
lidity of this presumption. As we have dem-
onstrated, there is extensive mobility be-
tween "male" and "female" job categories.
This in itself is inconsistent with Polachek's
model. But it also suggests that the use of
current occupation as a proxy for occupa-
tional history is inappropriate and may pro-
vide misleading information about whether
job choice is conditioned by expectations
about future work or whether experience
garnered in "female" jobs results in lower
wage growth and less depreciation than does
· ~ . ~~ ,, .
experience garnered in ma he Jo as.
We used the longitudinal nature of the
PSID to develop more direct tests of the
following two predictions of the human cap-
ital model:
1. Wage growth and depreciation are
lower for work experience gathered in "fe-
male'? jobs than in "male" jobs.
2. Women workers with extensive time
out and frequent interruptions are more likely
to have concentrated their work experience
~~ r i,
in tema he Jo as.
We find virtually no support for the predic-
tions.
To test the first proposition, we modified
our basic wage-growth equation to include
the sex typing of experience. In Eq. (7) we
did not distinguish whether years in the en
segment involved work in male-dominated
or female-dominated jobs. Since both in-
dustry and occupation were reported for 9
of the 13 years under investigation, we could
classify years in er that involved work in
"female" and "male" jobs.2i We combined
this with the information on work hours to
create four new variables:
1. the number of years of en, that were
full-time in male-dominated jobs (e~-fuit-m~,
2. the numbers of years of er that were
part-time in male-dominated jobs (e~-part-
3. the number of years of en that were
full-time in female-dominated jobs (e~-fuil-
fct), and
4. the numbers of years of el that were
part-time in female-dominated jobs (e~-part-
fd).
This decomposition of eL yields the following
equation:
bind= o~~l~/\s + a~Oe*-part +
o`~e*-fuii +
ar,4eL,-full-md +
aL~5eL,-part-md +
16eL,-fUi[-fti +
cx~7e~-part-fcl.
(8)
As with the basic measures of en, these four
new variables are obtained in each of the
annual interviews and do not rely on ret-
rospective reports by either women workers
or their husbands. These variables also pro-
vide more complete measures of the extent
to which work experience was acquired in
"female" jobs than do measures of occupa-
tion that are taken at a single point in time.
The results of estimating Eq. (8), shown
in Table 10-3, columns 5 and 6, provide little
support for the argument that wage growth
is much higher for male-dominated work than
for female-dominated work, especially for
white women. A much more important fac-
tor was whether the work performed in a
2} Industry is coded into 2-digit categories for 1971
to 1979. Occupation is coded into 1-digit census cat-
egories for the years 1971 to 1974 and into 2-digit cat-
egories for all the years thereafter. For each occupa-
tion-industry subgroup, we calculated a measure of
percent female. If there were more than 50 percent
women in that subgroup, we called it a "female-dom-
inated" job. (See Corcoran et al., in press, for a more
complete description of this procedure.) Since we only
have measures of occupation and industry for the last
9 years of the study, we do not break experience in e.
(which tended to occur early on in 1967 to 1979) into
"male" and "female" components.
OCR for page 188
188
MARY CORCORAN, GREG I. DUNCAN, AND MICHAEL PONZA
particular kind of job was part time or full
time. For white women a year of full-time,
male-dominated work was associated with a
2.6 percent increase in hourly wages, while
a year of full-time, female-dominated work
was associated with a 1.9 percent increase
in hourly wages. The differences between
these two coefficients were not significant.
For black women a year of female-domi-
nated, full-time work was associated with a
2.1 percent increase in wages. This com-
pares to a 0.8 percent increase for a year of
fi~-time, male-dominated work. Again, the
difference was not significant. Part-time work
experience, whether in "male" or "female"
jobs, had no significant effects on wages for
either blacks or whites.
We examined whether the sex typing of
women's work experience affected the rate
of depreciation during labor force withdraw-
als by interacting the two labor force with-
drawal measures (h* and had with a measure
of the average percent female in each wom-
an's occupation-industry combination over
the 13-year period. These interaction terms
were always insignificant when added to the
wage-change Eq. (81. This suggests that de-
preciation does not differ for "male" and
`d ~ ,, .
tema e Jo as.
The 13 years of PSID data allow a direct
test of the second proposition that work-
ers who expect discontinuous labor force ca-
reers will concentrate in "female" jobs. If
this hypothesis is correct, then the sex typ-
ing of work experience over the years 1967
to 1979 ought to be positively related to time
out of the labor force in 1967 to 1979, in-
termittency of work participation in 1967 to
1979 (measured by number of labor force
withdrawals), and whether working in 1979.
Results of this exercise (see Corcoran et al.,
in press) confirm England's finding of no
relationship between discontinuity of work
22 The .024~3 coefficient estimated for years of part-
time, male-dominated work for black women appears
to be out of line with the other results. This estimate
is based on a small number of observations, as reflected
in its large standard error.
and sex typing of concurrent occupation.
None of these three measures of labor force
discontinuity over 1967 to 1979 had a sig-
nif~cant, positive relation to the sex typing
of work experience over that period.
SUMMARY AND CONCLUSIONS
The wage change models of Mincer and
Ofek (1982) and those from our own work
yield similar results. Women who drop out
of the labor force have lower real wages when
they return to work than they had when they
left work. However, the period following
the return is characterized by rapid wage
growth, and the net loss in wages from drop-
ping out is small. This result reconciles the
apparently contradictory findings from cross-
sectional studies about the size of the de-
preciation effect, because cross-sectional
analyses pick up both short-run and long-
run depreciation effects. Short-run effects
are likely given more weight in an analysis
of women aged 30 to 44 (a group likely to
have recently completed labor force with-
drawals) than in analyses run on women in
a broader age range.
How does this empirical evidence affect
our understanding of the process of female
wage determination? The observed wage loss
and rebound pattern is certainly consistent
with the Mincer-Ofek story of human capital
depreciation and restoration. This pattern is
consistent with other stories as well- the
job mismatch argument and the probation-
ary period argument. We do not have the
necessary data to disentangle these argu-
ments. Regardless of the reason, the rapid
rebound of wage losses after labor force
withdrawals means that the wage losses as-
sociated with these withdrawals cannot ex-
plain much of the male/female wage gap.23
23 The fact that women work fewer years and more
part-time years than men work does explain a substan-
tial (one-third to two-fifths) part of the wage gap be-
tween men and women workers in a broad age range
(Corcoran and Duncan, 1979~. Note, however, that the
bulk of the sex-based wage gap differences is still unex-
plained by male/female work history differences.
OCR for page 189
WORK EXPERIENCE, TOB SEGREGATION, AND WAGES
189
Women are often urged to choose part-
time work rather than to stop work alto-
gether to keep their "hands in." Our results
provide little evidence that the wage con-
sequences of these two alternatives differ.
Part-time work experience is not rewarded
for women particularly for white women.
And the long-run wage penalties due to la-
bor force withdrawals are small. The deci-
sion whether to work full time or part time
is considerably more important than is the
choice between part-time work or no work.
These part-time results, like the wage loss
and rebound results, are consistent with
several quite different labor market scena-
rios a human capital mode! of lower train-
ing during part-time work, an imperfect in-
formation model, or an institutional model.
Again, we do not have the necessary data
to disentangle the competing explanations,
since they each involve a different interpre-
tation of the same employer behavior.
We also investigated the human capital
models that explain job segregation as the
result of women's discontinuous work his-
tory patterns. Such models emphasize two
costs of discontinuous work participation:
depreciation and forgone wage growth. The
rapid restoration of wage losses in the period
immediately following labor force withdraw-
als suggests that the first cost might be quite
small. For these human capital explanations
to hold, three things must occur: (1) there
should be considerable immobility between
"male" and "female" job sectors, (2) wage
growth and depreciation should be lower for
work in "female" jobs than for work in "male"
jobs, and (3) women with discontinuous work
careers will be more likely to choose "fe-
male" jobs than will women with continuous
work careers.
We find little evidence for any of these
propositions. First, there is considerable
mobility between "male" and "female" job
types. We did not find that either wage
growth or depreciation varied significantly
with the sex typing of work experience. These
results are consistent with England's cross-
sectional work. Finally, women with dis-
continuous work careers were no more likely
to have worked at "female" jobs than were
women with more continuous work expe-
rience.
These results also have implications for
models of job segregation other than the
human capital model. Many models of job
segregation either implicitly or explicitly as-
sume that there is rigid segmentation be-
tween "male" and "female" job sectors and
that there are fewer promotion and/or train-
ing opportunities in the "female" job sector
than in the "male" job sector. The analyses
reviewed in this paper suggest these as-
sumptions are likely wrong.
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Representative terms from entire chapter:
human capital
