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, [email protected]. 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 References Angrist, Joshua D. and Jörn-Steffen Pischke. 2009. Mostly harmless econometrics: an empiricist's companion. Princeton University Press. Barcellos, Silvia Helena. 2010. “Legalization and the economic status of immigrants.” RAND Working Paper Series, no. WR-754. Borjas, George J. and Marta Tienda. 1993. “The employment and wages of legalized immigrants.” International Migration Review 27, 712-47. Chi, Miao. 2015. “Does Intermarriage Promote Economic Assimilation among Immigrants in the United States?” International Journal of Manpower, forthcoming. Chi, Miao and Scott Drewianka. 2014. “How Much is a Green Card Worth? 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Gregory. 2005. “Intermarriage and the Economic Assimilation of Immigrants.” Journal of Labor Economics 23, 135-174. Meng, Xin, and Dominique Meurs. 2009. “Intermarriage, language, and economic assimilation process: a case study of France.” International Journal of Manpower 30, 127144. Pan, Ying. 2012. “The impact of legal status on immigrants’ earnings and human capital: evidence from the IRCA 1986.” Journal of Labor Research 33, 119-142. Passel, Jeffrey S., D’Vera Cohn, and Molly Rohal. 2014. “Unauthorized immigrant totals rise in 7 States, fall in 14.” Washington D.C.: Pew Research Center. Passel, Jeffrey S., Mark H. Lopez, D’Vera Cohn, and Molly Rohal. 2014. “As growth stalls, unauthorized immigrant population becomes more settled.” Washington D.C.: Pew Research Center. Ruggles, Steven, Matthew Sobek, Trent Alexander, Catherine A. Fitch, Ronald Goeken, Patricia K. Hall, Miriam King, and Chad Ronnander. 2008. Integrated Public Use Microdata Series: Version 4.0. Minneapolis, MN: Minnesota Population Center. Stock, James H., and Morohiro Yogo. 2002. “Testing for weak instruments in linear IV regression.” NBER Working Paper Series, Technical Working Paper 284. Wooldridge, Jeffrey W. 2002. Econometric Analysis of Cross Section and Panel Data. Cambridge, MA: MIT Press. 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
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