1. law, govt and politics
2. immigration

Improved Legal Status As the Major Source of Earnings Premiums Associated
with Intermarriage: Evidence from the 1986 IRCA Amnesty
Miao Chi*
Abstract: Exploiting a natural experiment, this paper uses the 1990 U.S. Census data and the
1986 Immigration Reform and Control Act (IRCA) amnesty to investigate the major
mechanism through which intermarriage influences immigrants’ earnings. My strategy
involves comparing international marriage premiums received by two groups of Mexican
immigrants who arrived before and after the cutoff date of eligibility. Both groups face similar
language and culture related obstacles and have to adapt themselves to the new environment,
except that unauthorized Mexican workers who arrived before 1982 could obtain legal status
through the amnesty while those who arrived after the cutoff date obtained legal status through
marriage to a U.S. citizen. Instrumental Variables estimates show a significantly larger
intermarriage premium for Mexican immigrants who migrated after the cutoff date and no
statistically significant intermarriage premium is found in the pre-1982 group. The 35 percent
premium gap indicates a large effect of intermarriage on immigrants’ labor market outcomes,
operating primarily through an improvement of legal status.
JEL Classifications:
J61, J12
Keywords: immigration, legal status, economic assimilation, international intermarriage
*
Drew University, Dept. of Economics and Business Studies, 36 Madison Ave., Madison, NJ 07940, mchi@drew.edu.
Tel.: 973-408-3833. Fax: 973-408-3142.
I. Introduction
As one of the most popular destinations for international migrants, the United States has
a large immigrant population consisting of both legal and unauthorized immigrants who either
entered the country without inspection or overstayed their visa. According to the Pew Hispanic
Center, the size of the unauthorized population was estimated to be 11.2 million in 2012,
including 5.85 million from Mexico alone (Passel, Cohn, and Rohal 2014). The unauthorized
immigrant population had risen sharply before plunging between 2007 and 2009. The number
has stabilized since the end of the Great Recession, and those who remain are more likely to
be long-term residents with U.S.-born children (Passel et al. 2014). For most immigrants, the
obtaining of work permits, permanent resident status, and eventually citizenship is a significant
leap forward and can be of great value to their economic assimilation and social incorporation.
This is especially true for undocumented workers as illegal status greatly limits the types of
jobs they are able to obtain, which generates inefficiency in the labor market while preventing
immigrants from working on jobs that best match their skills.
Using the 1990 U.S. Census data, this paper exploits the last general amnesty for
unauthorized workers to disentangle the main mechanism through which international
intermarriage improves immigrant earnings. As an attempt to curb illegal immigration,
Congress passed the bill known as the Immigration Reform and Control Act (IRCA) in October
1986, which granted permanent legal status to 2.7 million unauthorized immigrants. A great
majority of the amnesty beneficiaries were from Mexico (over 2.2 million) and Central
America (over 0.3 million). My strategy is to take the difference between the estimated
intermarriage premium received by the post-1982 group of Mexican-born workers and the pre-
1
1982 Mexican immigrants used as a control/placebo group, which has the same rationale
behind a difference-in-difference-in-differences strategy. Since intermarriage is likely
endogenous, I will use appropriate instruments.
As the IRCA essentially granted legal status to all unauthorized Mexican immigrants who
entered the country before January 1, 1982, intermarried unauthorized immigrants in the post82 group would receive an additional benefit of obtaining permanent legal status compared to
their counterparts in the pre-82 group who obtained legal status outside of marriage. On one
hand, a significant increase in the size of the intermarriage premium received by those not
eligible for the amnesty, over and above that received by immigrants who obtained legal status
through the amnesty, would suggest that improvement of legal status is a major mechanism
through which intermarriage influences an immigrant’s labor market outcomes and is of great
value in an immigrant’s economic assimilation process. On the other hand, similar
intermarriage premiums between the two groups would indicate a greater importance of other
mechanisms such as language acquisition, cultural assimilation, or an access to extended social
networks and knowledge of local labor markets.
Intermarriage, defined as a marital union between a foreign-born and a native-born
individual, is viewed as both a measure of social assimilation and a factor producing it
(Lieberson and Waters 1986). Only a few economics papers have estimated the causal effects
of intermarriage on immigrants’ earnings. Meng and Gregory (2005) find a positive
intermarriage premium of 15 to 23 percent among immigrants in Australia, and Meng and
Meurs (2009) find a 25 to 35 percent premium in France. Kantarevic (2004) finds no
intermarriage premium in the 1970 and 1980 U.S. Censuses once selection bias is accounted
2
for, but Chi (2015) finds a 4 to 6 percent premium in the 2000 U.S. Census, with larger
premiums for Latinos and some other ethnic groups. Furtado and Theodoropoulos (2009) also
find that marriage to a U.S. native increases the probability that an immigrant is employed.
Much less is known about the specific mechanisms through which intermarriage
influences an immigrant’s labor market outcomes, although several have been suggested. One
possibility is language acquisition. Several studies suggest that immigrants with better
language skills assimilate much faster (Chiswick and Miller 1995), and having a native-born
spouse may help to improve their English skills. Intermarriage may also encourage cultural
assimilation and provide access to social networks. For example, an immigrant could obtain
specific knowledge of employment opportunities or local labor market institutions. Furtado
and Theodoropoulos (2010) examine the effect of intermarriage on an immigrant’s employment
status and suggest that the returns to marrying a native arise partially from networks acquired
through marriage. Another possible benefit associated with intermarriage is the improvement
of an immigrant’s legal status. With permanent resident status, an immigrant is able to work
full-time legally and enjoys higher earnings than his counterparts who are either not allowed
to work or turned down by potential employers due to their alien status. Chi and Drewianka
(2014) find permanent access to labor markets to be a significant source of earnings premiums
associated with marriage to a native.
This paper contributes to the very small literature on the mechanisms through which
intermarriage benefits an immigrant’s economic outcomes by exploiting the event of the 1986
IRCA amnesty to isolate the effect of improvement on legal status obtained through
intermarriage.
3
Most previous studies find positive effects of IRCA on wages of the affected immigrants;
the typical wage increase is estimated to be less than ten percent (Borjas and Tienda, 1993;
Cobb-Clark, Shiells, and Lowell, 1995; Kossoudji and Cobb-Clark, 2002; Barcellos, 2010; Pan,
2012). However, Rivera-Batiz (1999) and Lozano and Sorensen (2011) estimate much larger
wage gains associated with the amnesty program, around 15-20 log points for the average man
from Mexico. While most of earlier studies use small datasets, the major challenge to empirical
studies of the effects of legalization on the immigrant economic outcomes remains as the lack
of information about legal status on national surveys that are large enough to sufficiently
represent the immigrant population. This paper examines the Census data to reflect the wage
gain associated with legalization.
The results will demonstrate a statistically significant intermarriage premium difference
of 35 percent between the two groups of Mexican immigrants. The larger and significant
intermarriage premium received by those who entered after the cutoff reflects a large wage
effect of intermarriage through legalization. The results are very robust to alternative
specifications, sample selection criteria, and choices of excluded variables.
II. Model Specification
My baseline model is as follows:
Yijg   g  X ij ' 1  Yj 2  Nijg  g  Dijg g   ijg
where the dependent variable 𝑌𝑖𝑗𝑔 is the log hourly earnings of Mexican immigrant i who
immigrated during period g and is now living in place j. The vector 𝑋𝑖𝑗 contains quadratic terms
in the immigrant’s age and the reconstructed number of years since he immigrated, and dummy
4
variables indicating educational attainment, English proficiency, school attendance, residence
in metro areas, regions, disability and veteran status. 1 𝑌𝑗 is the average log hourly earnings of
native men with similar educational attainment in place j, which I include to control for
differences in price levels across cities within the region.
𝑔
The remaining explanatory variables are the main object of interest. 𝑁𝑖𝑗 is a dummy
𝑔
variable that equals 1 if the immigrant is married to another immigrant, and 𝐷𝑖𝑗 equals 1 if the
immigrant is married to a native. Thus, if I denote the Mexican men who immigrated before
and after January 1, 1982 in my sample respectively as g=pre and g=post, the intermarriage
premium is (𝛿𝑝𝑟𝑒 − 𝛾𝑝𝑟𝑒 ) for Mexicans who entered before the cutoff date and (𝛿𝑃𝑜𝑠𝑡 − 𝛾𝑝𝑜𝑠𝑡 )
for Mexicans who entered after the cutoff date, and I attribute the difference (𝛿𝑝𝑜𝑠𝑡 − 𝛾𝑝𝑜𝑠𝑡 −
𝛿𝑝𝑟𝑒 + 𝛾𝑝𝑟𝑒 ) to the fact that the unauthorized Mexican immigrants who immigrated in 1982 or
after may gain legal status through intermarriage while those who entered before 1982 already
have that legal status through the amnesty. This difference would then reflect the effect of
intermarriage on immigrants’ earnings due to improved legal status.
One’s choice between marrying a native and marrying another immigrant is likely
endogenous and might be determined by unobserved characteristics such as ability, diligence,
motivation, and open-mindedness, and these characteristics also influence one’s earnings.
Another issue may be that workers who have more to gain from legal status may pursue it more
aggressively, e.g., by directed search for a native spouse. To address the endogeneity concern,
I rely heavily on the inclusion of appropriate control variables and the use of geography-based
instrumental variables that generate independent variation in marital status.
1
The disability dummy equals one if the individual has a lasting physical or mental health condition that causes
difficulty working.
5
The standard method to address endogeneity is to instrument for the marriage dummies,
which requires variables Z that predict the variation in marital status but are otherwise unrelated
to individuals’ earnings. An additional complication arises here because the endogenous
explanatory variables are dummy variables. While in principle one could use Z directly as
instruments for immigrants’ marital statuses, which would essentially amount to using a linear
probability model in the first-stage IV regression, it is generally more efficient instead to use
those factors as excluded exogenous variables in a non-linear model predicting marital status,
and then to use the fitted values from those non-linear models as instruments in a standard IV
procedure.2 I implement this approach using a multinomial logit to predict immigrants’ marital
status. The same “efficient instrumental variables” procedure is used in Chi and Drewianka
(2014), which contains the more extensive explanation of the approach.
The excluded exogenous variables considered here all involve the demographic
composition of the population across locations. Similar strategies have been used in most
previous work on intermarriage (Kantarevic, 2004; Meng and Gregory, 2005; Furtado and
Theodoropoulos, 2009; Meng and Meurs, 2009; Chi and Drewianka, 2014). My preferred
specification follows Meng and Gregory’s study and uses (1) the share of local unmarried
women who are Hispanic immigrants, and (2) the sex ratio among unmarried Hispanic
immigrants.
The first variable reflects the availability of potential foreign-born Hispanic spouse and I
anticipate that it will be a good predictor for whether a married immigrant man has a foreignborn or native-born wife. The sex ratio variable is intended to reflect the competition among
2
For a more general discussion of this approach, see Wooldridge (2002, 230-237) or Angrist and Pischke
(2009, 190-192).
6
fellow Mexican immigrant men. A very unbalanced sex ratio will reduce the overall marriage
rate, so I expect it to be most useful for predicting whether an individual is married or single.
Table 4 will show that the estimated difference between the post- versus pre-1982 Mexican
intermarriage premiums is quite similar if I use alternative excluded exogenous variables. I
also consider Kantarevic’s (2004) instrument, the ratio of immigrants to natives among local
unmarried women, relative to the same ratio throughout the entire U.S. Five additional
variables are considered as well: the ratio of immigrants to natives within the location’s
population of unmarried Hispanic women, the share of local women who are unmarried, the
share of local women who are Hispanic (including but not limited to immigrant women), the
sex ratio among all local unmarried people, and the sex ratio among all local people.
III. Data and Descriptive Statistics
My data is from the 5 percent sample of the 1990 U.S. Census, which was obtained
through the Integrated Public Use Microdata Series (Ruggles et al., 2008). The Census data
provides a large sample size and captures a large share of the unauthorized population as
suggested by the changes in the population between census years that cannot be explained by
natural population growth or authorized immigration (Lozano and Sorensen, 2011). I restrict
the sample to Mexican-born men aged 16 to 44, who worked at least one week in 1989 and
do not reside in group quarters. I drop observations with missing data on the man’s earnings,
his (and if applicable, his wife’s) birthplace, or his year of immigration.
It is not possible to determine the age at which respondents married in the 1990 Census,
so no information is available for me to tell whether immigrants were married when they
7
arrived in the U.S. I also exclude a small number of men with non-Hispanic immigrant wives.
Intermarriage is defined as a marital union between a man born in Mexico and a wife
born in the U.S. or Puerto Rico (since all Puerto Ricans are U.S. citizens). If I exclude
Mexicans who married Puerto Rican-born wives, the resulting estimates are indistinguishable
from those reported in Table 3.
The dependent variable is the natural log of individuals’ pre-tax wage and salary
income per hour worked in 1989, including wages, salaries, commissions, cash bonuses, tips,
and other money income received from an employer.
Key controls include dummies for English proficiency (=1 if the person speaks English
well or better) and each level of education. 3 The 1990 Census reported respondents’ year of
immigration as ranges rather than the exact year in which the immigrant entered the U.S. I
thus reconstruct a continuous years since migration variable using the midpoint in each
range. 4 Local demographic variables, including all instruments, are computed at the level of
metropolitan areas. For each state, I have also defined a separate “at-large” market consisting
of everyone living outside a metro area.
The resulting sample includes 57,946 men born in Mexico. Table 1 presents summary
statistics. A larger share of the pre-82 group is married (67 percent) compared to the post-82
group (35 percent), which might be explained by the average age difference of four years
between the pre-82 and post-82 groups. Among those married, 25% of the pre-82 group is
married to a native spouse, and the intermarriage rate for the post-82 group is 18%. Given
3
It is standard in the literature to use a binary variable to control for English ability. The 1990 Census collected
information regarding each respondent’s highest degree or level of school completed instead of years of schooling.
4
Specifically, the reconstructed years since migration variable is equal to 1.5, 4.5, 7, 8, 13, 18, 23, 28, 35.5 and
46 corresponding to the following periods of entry: 1987-1990, 85-86, 82-84, 80-81, 75-79, 70-74, 65-69, 60-64,
50-59, 1949 or earlier.
8
the age difference between the two groups, it is not surprising that the pre-82 group earns
more on average and has greater English proficiency than those who immigrated in 1982 or
after. Intermarried men in both groups have more education, greater English proficiency, and
slightly younger than those married to other Hispanic immigrants. The raw intermarriage
premium for Mexicans who immigrated before 1982 is slightly larger than for the post-82
group (7 percent versus 4 percent). However, note that post-1982 men are much less
positively selected into intermarriage on the basis of education and English proficiency, two
of the observable characteristics most associated with positive earnings premiums. This
difference in selection into intermarriage suggests that the difference in the raw intermarriage
premiums will understate the causal effect of legal status on wages, and the results in the next
section suggest that the post-1982 immigrants are also negatively selected into intermarriage
on the basis of unobservables. This is consistent with the idea that people who would benefit
most from legal status were also more likely than others to search for a native spouse.
[Table 1 about here.]
IV. Empirical Results
A. Determinants of Intermarriage
Table 2 presents results from a multinomial logit predicting the probability that a Mexican
immigrant is single, married to another Hispanic immigrant, or married to a native. It reports
the coefficients of key determinants when the pre-82 and post-82 groups are estimated
separately. This specification is used in all later estimations to account for the possibility that
9
the two groups might have different incentive to marry a U.S. citizen and thus selected into
intermarriage differently.
[Table 2 about here.]
Both excluded variables that are used to obtain the nonlinear fitted values of the
intermarriage and non-intermarriage dummies are strong predictors of immigrants’ marital
statuses. As expected, the share of local unmarried women who are Hispanic immigrants is a
statistically significant predictor of both intermarriage and non-intermarriage, for both pre- and
post-1982 immigrants. In particular, greater availability of potential foreign-born Hispanic
spouses reduces the probability of intermarriage and increases the probability of marriage to
another Hispanic immigrant. Being single is the base category. The sex ratio is also a significant
predictor of both marriage types: a balanced sex ratio increases the probability of being married
versus staying single.
While my excluded variables are statistically significant predictors of marriage and
intermarriage for both the pre- and post-1982 immigrants, the coefficient on the variable that I
expect to predict intermarriage (in the first row) is substantially larger for the pre-1982
immigrants. It suggests that the post-1982 immigrants’ decision to marry a native is less
sensitive to their relative supply, which is consistent with the notion that they are especially
interested in marrying such women regardless of their availability. It is also noteworthy that
the coefficient on this variable in the non-intermarried equation is positive for the pre-1982
immigrants but negative for the post-1982 immigrants. This suggests that the pre-1982 group
is more likely to marry when there are a lot of immigrant women around while the post-1982
group might rather not marry another immigrant, such as if there were some strong reason (e.g.,
10
a green card) that they are holding out for a native wife.
Immigrants who reside in places where the average earnings of native men with similar
educational attainment are higher are less likely to intermarry. The reason behind this negative
correlation might be that the higher average earnings of native men may reflect location
characteristics such as size, popularity, economic opportunities, and living expenses. Large
places with higher average earnings levels may attract more immigrants than small cities do,
and this may increase the pool of marriageable Hispanic immigrant women. Also, immigrants
in these places might face more competition from native-born men in the marriage market.
Immigrants proficient in English are more likely to marry a native.
B. Main Results
Table 3 presents the main results: estimates of the intermarriage premiums for the pre- and
post-1982 groups, as well as the difference between them. The first column shows the estimates
from OLS estimation of the wage equation, with intercepts and the effects of years since
migration allowed to differ between the pre- and post-1982 groups. Both groups receive
statistically significant intermarriage premiums, over and above the premium received by
immigrants married to another Hispanic immigrant: 3.2 percent for those who immigrated
before 1982 and 6.2 percent for the post-1982 group, though the difference is statistically
insignificant.
[Table 3 about here.]
The second column reports IV estimates; recall that the instruments are predicted
11
probabilities of marriage and intermarriage from multinomial logits. 5 The instruments for the
pre- and post-82 groups are calculated from separate logits on the two groups. The lower panel
reports statistics from the first-stage regressions. Since the smallest of the reported F-statistics
is 153, roughly 15 times larger than the minimum guideline suggested by Stock and Yogo
(2002), I conclude that the multinomial logit fitted values are strong instruments for the two
marital status dummies in equation (1).
Estimates of the two groups’ intermarriage premiums are very different: an insignificant
9.4 percent for Mexicans who were eligible for the amnesty and a large and statistically
significant 44.8 percent for those who entered after the cutoff date and thus did not qualify for
the amnesty. The difference in the intermarriage premiums between the two groups is a
statistically significant 35.4 percent, which reflects the main mechanism through which
intermarriage benefits immigrants’ earnings.
C. Robustness
Table 4 reports estimates using alternate sample selection criteria and excluded
exogenous variables. Estimations reported in the top panel are conducted to make treatment
and control groups as comparable as possible. The first column presents IV estimates from a
sample that includes only Mexican men who immigrated after 1975 since some people in the
pre-82 group migrated decades ago and might not be relevant as a control group. The preand post-1982 samples both cover about 7 years here. The estimates imply that intermarried
Mexican immigrants who are not eligible for the amnesty earn a 29.3 percent premium, but
5
Results are very similar if I use fitted values from a multinomial probit rather than a logit as instruments.
12
the estimate for the pre-1982 group is a much smaller 6.7 percent and statistically
insignificant, so the estimated post-pre difference is about 23 percent. Given that the post1982 group is about 4 years younger than the pre-1982 group, the next column uses only
immigrants aged 25 or above to reduce the number of young Mexicans who might not be
very comparable to those in the pre-1982 group; much as in Table 3, the estimated post-pre
premium difference is around 35 percent and statistically significant. Estimates are very
similar if I exclude men from the top three cities with largest Mexican population (column
3). Since over half of unauthorized Mexican workers are high school dropouts, results should
still hold if I examine Mexicans with lower educational attainment and thus more likely to
be unauthorized. Column 4 includes only high school dropouts and the resulted post-pre
difference is 37 percent, significant at the one percent level.
The lower panel reports results using the alternate variables described at the end of
Section II. The estimated post-pre premium difference is remarkably similar to both one
another and to those from Table 3: always close to 36 percent and statistically significant at
the one percent level. For Mexicans who migrated after the cutoff date, the estimated
intermarriage premium is almost always near 45 percent and significant, but for the pre-1982
group the estimated intermarriage premiums are never statistically significant.
[Table 4 about here.]
D. Additional Control Groups
Since the 1986 IRCA amnesty programs mainly affected Mexicans and Central
Americans, we should not find significant post-pre intermarriage premium difference among
13
groups not affected by the amnesty. Table 5 reports the estimated intermarriage premiums for
the pre- and post-1982 groups among South American- and Puerto Rican-born men. Column
1 repeats the result for Mexicans from Table 3 for ease of comparison. The middle column
reports results for South Americans. In contrast to Mexicans, both pre- and post-82 South
American groups receive a significant intermarriage premium (52 and 40 percent,
respectively), suggesting that immigrants not impacted by the amnesty could benefit much
from legal status obtained through intermarriage. It is reassuring that the post-pre difference
is statistically insignificant. The last column lists the results using Puerto Rican-born men.
Since Puerto Ricans are U.S. citizens by birth, it is encouraging but not surprising that neither
group receives a large or statistically significant intermarriage premium.
[Table 5 about here.]
V. Conclusion
While marriage assimilation of immigrants in the host country has attracted some research
effort, little is known about the specific mechanisms through which intermarriage influences
an immigrant’s labor market outcomes. This article uses a novel approach to evaluate the wage
gain associated with legalization for immigrants in the U.S.; it explores the possibility of
improved legal status as a major mechanism through which intermarriage facilitates the
immigrant economic assimilation process.
Using the 1990 U.S. census data and the cutoff eligibility date of the 1986 IRCA amnesty,
this paper focuses on immigrant men born in Mexico and finds that Mexicans not eligible for
the amnesty receive a positive earnings premium associated with marrying a native while no
14
evidence of intermarriage premium is found for Mexicans eligible to receive permanent legal
status through the amnesty. The size of the intermarriage premium received by the ineligible
group of Mexicans is around 45 percent, leading to a statistically significant premium
difference of 35.4 percent between the control and treatment groups, equivalent to 4,974 dollars
in the year 1989.
There are many possible explanations for this positive relationship between intermarriage
and earnings for Mexican immigrants in the control group. For example, language and cultural
assimilation, expansion of social networks, or legal benefits such as obtaining a green card or
citizenship. By comparing Mexican immigrants who entered after the cutoff point with their
counterparts in the treatment group who could obtain legal status through the amnesty program,
this paper is able to disentangle the specific mechanism through which intermarriage affects an
immigrant’s earnings. The finding of significantly larger intermarriage premiums for Mexican
immigrants in the control group suggests that legal benefits might be the major source of
intermarriage premiums for immigrants in the U.S. and are of great value to immigrants in the
U.S. labor market. The sizable wage gain is not surprising since job mobility and a significant
increase in bargaining power might be the main channel through which legal status benefits
unauthorized workers.
Given the recent development in immigration reform and heated debate on immigration
policy regarding legalization, my finding of a large wage gain from legal status also provides
supportive evidence for President Obama’s recent executive order on immigration. As the
targeted beneficiaries of the executive order, unauthorized immigrants with U.S.-born children
or children with Permanent Resident status are more likely to stay in the U.S. for a sustained
15
period of time and would benefit greatly from being able to work in better matching
occupations without the fear of deportation. Issuing work permits to this group of immigrants
would effectively increase their job mobility and bargaining power, thus achieving the goals of
improving their labor market outcomes, all accomplished without the use of amnesty programs
that might cause much stronger oppositions.
16
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17
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18
Table 1: Summary Statistics, Single and Married Immigrant Men from Mexico Aged 16-44
Log Hourly
Age
Mean
SD
Speaks
English
"well" or
better
Mean SD
27.00
33.54
33.77
32.86
0.71
0.64
0.58
0.83
Earnings
Annual Earnings
Mean
SD
Mean
SD
12,308
18,544
17,939
20,345
1.85
2.11
2.09
2.16
0.63
0.63
0.62
0.66
Educational Attainment
High
Sch.
Dropouts
High Sch.
Graduates
Some
College
College
Graduates
64.4%
19.0%
13.9%
2.7%
11,937
73.9%
78.4%
60.9%
12.8%
11.2%
17.3%
10.4%
8.3%
16.7%
2.9%
2.1%
5.2%
23,914
17,825(72%)
6,089(25%)
Number of
Obs.
Pre 1982:
Single Men
All Married Men
Married to a Hispanic Immigrant
Married to a Native
10,109
13,072
12,177
15,290
6.70
5.83
5.76
5.99
0.45
0.48
0.49
0.37
Post 1982:
Single Men
9,305 7,119
1.66 0.57 23.25 4.60
0.35 0.48
75.2%
14.9%
8.1%
1.8%
14,344
All Married Men
12,586 10,156
1.83 0.62 28.48 5.76
0.41 0.49
71.7%
12.7%
10.1%
5.6%
7,751
Married to a Hispanic Immigrant
12,449 10,241
1.82 0.62 28.73 5.82
0.35 0.48
72.3%
12.1%
9.2%
5.4%
6,355(82%)
Married to a Native
13,211 9,739
1.86 0.62 27.30 5.29
0.67 0.47
64.3%
15.1%
13.8%
6.8%
1,396(18%)
Note: Data are from the Public Use Microsample of the 1990 U.S. Census of Population. All calculations use person-level sample weights. The sample excludes men
who are divorced, widowed, married to a non-Hispanic immigrant, or married to an absent spouse.
19
Table 2: Determinants of Probabilities of Marital Outcomes with Respect to
Selected Explanatory Variables in Multinomial Logit Model
Intermarried
Non-intermarried
Explanatory Variable
Est.
SE
P
Share of local unmarried women who are Hispanic immigrants
Pre-1982
-0.705
0.080
0.00
Post-1982
-0.187
0.027
0.00
Sex ratio among local unmarried Hispanic immigrants
Pre-1982
0.021
0.020
0.29
Post-1982
0.034
0.009
0.00
Average wage of local native-born married men in educational group
Pre-1982
-0.255
0.024
0.00
Post-1982
-0.082
0.011
0.00
Dummy: English proficiency
Pre-1982
0.100
0.006
0.00
Post-1982
0.049
0.005
0.00
Test: Excluded variables are jointly
significant
Pre-1982
Post-1982
Est.
SE
P
0.419
-0.172
0.124
0.084
0.00
0.04
0.092
0.143
0.031
0.028
0.00
0.00
0.224
-0.068
0.054
0.048
0.00
0.15
-0.072
-0.018
0.010
0.007
0.00
0.00
2(2)
P
2(2)
P
64.3
0.00
18.5
0.00
65.4
0.00
31.1
0.00
Pseudo R2
Pre-1982
Post-1982
No. of Observations
0.19
0.20
57,946
Notes: The reported estimates are computed by evaluating the marginal effect of the individual
variable for each observation and then averaging over the sample of these marginal effects. Being
single is the base category. The first two explanatory variables comprise my preferred set of
excluded exogenous variables. Other non-reported controls include the individual's age and its
squared term, the reconstructed years since migration and its squared term, and dummy variables
indicating his English fluency, educational attainment, disability and veteran statuses, current
school attendance, urban residence, and nine geographic regions. The sample excludes men who
are divorced, widowed, married to a non-Hispanic immigrant, or married to an absent spouse. All
estimates use person-level sample weights, and standard errors correct for non-independence
(clustering) of observations within metropolitan areas.
20
Table 3: OLS and IV Estimates of the Wage Equation
OLS
Estimated Coefficients in Wage
Equation
Intermarriage Premium
Pre 1982:
Post 1982:
Post - Pre Difference
R2
Observations
IV
Est.
(x100)
SE
(x100)
P
3.2
6.2
3.0
1.2
2.0
2.3
0.00
0.00
0.18
Est.
(x100)
SE
(x100)
9.4
44.8
35.4
8.5
14.6
12.5
0.18
57,946
0.27
0.00
0.01
0.15
57,946
R2
0.19
0.42
0.14
0.36
First-stage IV regressions
Intermarriage - Pre 1982
Non-intermarriage - Pre 1982
Intermarriage - Post 1982
Non-intermarriage - Post 1982
P
F
406.3
1,421.8
152.7
1,090.6
P
0.00
0.00
0.00
0.00
Notes: The excluded variables used are (1) the share of local unmarried women who are Hispanic
immigrants, and (2) the sex ratio (women/men) among unmarried Hispanic immigrants. Other
controls included are the individual's age and its squared term, the reconstructed years since migration
and its squared term, and dummy variables indicating his English fluency, educational attainment,
disability and veteran statuses, current school attendance, urban residence, and nine geographic
regions. Intercepts and effects of years since migration and its squared term are allowed to differ
between those who migrated before 1982 and the post-82 group. The sample excludes men who are
divorced, widowed, married to a non-Hispanic immigrant, or married to an absent spouse. All
estimates use person-level sample weights, and standard errors correct for non-independence
(clustering) of observations within metropolitan areas.
21
Table 4: Selected IV Estimates of the Earnings Equation Using Alternative Sample Selection Criteria and Excluded Exogenous Variables
Sample:
Intermarriage Premium
Pre 1982:
Post 1982:
Post - Pre Difference
R2
Observations
Est.
SE
(x100)
(x100)
P
Exclude men who immigrated
before 1975
6.7
29.3
22.6
7.9
13.3
11.7
0.11
42,330
0.40
0.03
0.05
Est.
SE
(x100)
(x100)
P
Exclude men younger than
25
-1.3
33.3
34.6
10.8
16.6
12.1
Est.
SE
(x100)
(x100)
P
Exclude men from LA,
Houston, Chicago
0.90
0.05
0.00
11.9
36.9
25.1
0.07
39,267
12.7
17.3
14.3
0.35
0.03
0.08
0.15
31,807
Using Alternate Excluded Exogenous Variables
Intermarriage Premium
Pre 1982:
Post 1982:
Post - Pre Difference
R2
Observations
Excluded variables for
multinomial logit
8.0
44.0
36.1
8.8
14.6
13.1
0.15
57,946
0.36
0.00
0.01
Ratio of Hispanic
immigrants/natives among local
unmarried women / same ratio
for entire US
Sex ratio (women/men) among
all local people
9.0
43.3
34.3
10.6
15.2
13.6
0.16
57,946
0.40
0.00
0.01
Ratio of immigrants/ natives
among local unmarried
Hispanic women
Share of all local women
aged 16-44 who are
unmarried
Notes: Estimates use the same specification and methods as in Table 3.
21
8.5
45.9
37.4
9.1
15.5
13.2
0.13
57,946
0.35
0.00
0.01
Share of local native
unmarried women who are
of Hispanic descent
Sex ratio (women/men)
among all unmarried people
Est.
SE
(x100) (x100)
P
HS dropouts only
-6.5
30.5
37.0
8.6 0.45
14.8 0.04
13.0 0.00
0.12
41,708
Table 5: IV Estimates of the Intermarriage Premiums Received by Mexican, South
American, and Puerto Rican Men Aged 16-44
Mexicans
Est.
SE
(x100) (x100)
Intermarriage Premium
Pre 1982:
Post 1982:
Post - Pre Difference
R2
Observations
9.4
44.8
35.4
8.5
14.6
12.5
South Americans
P
0.27
0.00
0.01
Est.
SE
(x100) (x100)
52.2
40.1
-12.1
0.15
57,946
14.3
19.0
22.2
0.14
8,001
Notes: Estimates use the same specification and methods as in Table 3.
22
Puerto Ricans
P
Est.
(x100)
SE
(x100)
P
0.00
0.04
0.59
5.5
-13.9
-19.4
16.8
27.1
0.75
0.61
24.6
0.43
0.18
8,311