A recent article1 described increases in MMRs from 1999 to 2019, racial and ethnic disparities, and differences by US state. The use of bayesian models to provide more precise estimates of MMRs for small racial and ethnic subgroups is a strength, but not a panacea for the problem of small numbers. The utility of state-level estimates by subgroup is questionable when driven mainly by the priors and associated with wide uncertainty intervals, a concern for subgroups with fewer than 20 events each year. Additionally, failure to account for differential adoption of the pregnancy checkbox by US states over time biased the trend estimates and resulting conclusions of this study.1 | | The study authors acknowledged in the Discussion section that “it is possible that some of the increases in maternal mortality over time are due to an increasing number of states incorporating the pregnancy checkbox….”1 However, it is insufficient to simply note the effect of the pregnancy checkbox as a limitation, given numerous prior studies describing trends in maternal mortality in the context of changes in ascertainment. These studies have quantified the effect of the incremental adoption of the pregnancy checkbox on the standard certificate of death (which occurred from 2003 to 2017), showing that observed increases in MMRs from the early 2000s to 2017 are entirely or nearly entirely an artifact of changes in measurement over time, with no significant trends in MMRs once the checkbox was accounted for.2-4 It is highly likely that the increases reported in this study1 were an artifact of changes in ascertainment. | | Estimated racial and ethnic disparities in MMR trends are also subject to bias due to the incremental adoption of the checkbox by states over time, given the geographic concentration of specific subpopulations such as non-Hispanic American Indian or Alaska Native persons. The effect of checkbox implementation varies by age, race and ethnicity, state, and cause of death, with larger effects seen among people aged 35 years or older, among non-Hispanic Black individuals, and for nonspecific maternal causes of death.2-4 Consequently, it is important to explicitly account for the effect of the pregnancy checkbox and other changes in measurement5 over time when estimating trends and racial and ethnic disparities in MMRs as well as state-level patterns. Without accurate and comparable measurement of MMR trends and disparities, conclusions cannot be drawn about the effect of prevention efforts.

BACKGROUND AND OBJECTIVES: Although mortality rates are highest for infants of teens aged 15 to 19, no studies have examined the long-term trends by race and ethnicity, urbanicity, or maternal age. The objectives of this study were to examine trends and differences in mortality for infants of teens by race and ethnicity and urbanicity from 1996 to 2019 and estimate the contribution of changes in the maternal age distribution and maternal age-specific (infant) mortality rates (ASMRs) to differences in infant deaths in 1996 and 2019. METHODS: We used 1996 to 2019 period-linked birth and infant death data from the United States to assess biennial mortality rates per 1000 live births. Pairwise comparisons of rates were conducted using z test statistics and Joinpoint Regression was used to examine trends. Kitagawa decomposition analysis was used to estimate the proportion of change in infant deaths because of changes in the maternal age distribution and ASMRs. RESULTS: From 1996 to 2019, the mortality rate for infants of teens declined 16.7%, from 10.30 deaths per 1000 live births to 8.58. The decline was significant across racial and ethnic and urbanization subgroups; however, within rural counties, mortality rates did not change significantly for infants of Black or Hispanic teens. Changes in ASMRs accounted for 93.3% of the difference between 1996 and 2019 infant mortality rates, whereas changes in the maternal age distribution accounted for 6.7%. CONCLUSIONS: Additional research into the contextual factors in rural counties that are driving the lack of progress for infants of Black and Hispanic teens may help inform efforts to advance health equity.

Rural-urban differences in maternal mortality ratios (MMR) in the United States have been difficult to measure in recent years due to the incremental adoption of a pregnancy status checkbox on death certificates. Using 1999-2017 mortality and birth data, we examined the impact of the pregnancy checkbox on MMRs by rural-urban residence (large urban, medium/small urban, rural), using log-binomial regression models to predict trends as if all states had adopted the checkbox as of 1999. Implementation of the checkbox resulted in an average estimated increase of 7.5 maternal deaths per 100,000 live births (95% CI: 6.3, 8.8) in large urban areas (76% increase), 11.6 (95% CI: 9.6, 13.6) in medium/small urban areas (113% increase), and 16.6 (95% CI: 12.9, 20.3) in rural areas (107% increase), compared with MMRs prior to the checkbox. Assuming all states had the checkbox as of 1999, demographic-adjusted predicted MMRs increased in rural, declined in large urban, and did not change in medium/small urban areas. However, trends and urban-rural differences were substantially attenuated when analyses were limited to direct/specific causes of maternal death, which are likely subject to less misclassification. Accurate ascertainment of maternal deaths, particularly in rural areas, is important for reducing disparities in maternal mortality.

The COVID-19 pandemic has disproportionately affected Hispanic or Latino, non-Hispanic Black (Black), non-Hispanic American Indian or Alaska Native (AI/AN), and non-Hispanic Native Hawaiian or Other Pacific Islander (NH/PI) populations in the United States. These populations have experienced higher rates of infection and mortality compared with the non-Hispanic White (White) population (1-5) and greater excess mortality (i.e., the percentage increase in the number of persons who have died relative to the expected number of deaths for a given place and time) (6). A limitation of existing research on excess mortality among racial/ethnic minority groups has been the lack of adjustment for age and population change over time. This study assessed excess mortality incidence rates (IRs) (e.g., the number of excess deaths per 100,000 person-years) in the United States during December 29, 2019-January 2, 2021, by race/ethnicity and age group using data from the National Vital Statistics System. Among all assessed racial/ethnic groups (non-Hispanic Asian [Asian], AI/AN, Black, Hispanic, NH/PI, and White populations), excess mortality IRs were higher among persons aged ≥65 years (426.4 to 1033.5 excess deaths per 100,000 person-years) than among those aged 25-64 years (30.2 to 221.1) and those aged <25 years (-2.9 to 14.1). Among persons aged <65 years, Black and AI/AN populations had the highest excess mortality IRs. Among adults aged ≥65 years, Black and Hispanic persons experienced the highest excess mortality IRs of >1,000 excess deaths per 100,000 person-years. These findings could help guide more tailored public health messaging and mitigation efforts to reduce disparities in mortality associated with the COVID-19 pandemic in the United States,* by identifying the racial/ethnic groups and age groups with the highest excess mortality rates.

The coronavirus disease 2019 (COVID-19) pandemic has disproportionately affected racial and ethnic minority groups [1–5]. COVID-19 infection and mortality rates are higher among Hispanic/Latino, non-Hispanic Black, and non-Hispanic American Indian or Alaska Native (AI/AN) populations than non-Hispanic White populations [5]. Although most U.S. COVID-19 deaths have occurred among adults aged ≥65 years, younger persons represent a larger percentage of COVID-19 deaths in Hispanic/Latino, non-Hispanic Black, and non-Hispanic AI/AN populations [1]. These racial/ethnic groups also have younger age distributions across the population generally [3], and face increased risk of COVID-19 infection and related morbidity and mortality as a result of many different factors such as the degree of occupational exposure, housing or residential risk factors, the prevalence of preexisting health conditions, reduced access to care, and structural racism [1], [2], [3], [4].

Estimates of excess deaths, defined as the number of persons who have died from all causes, above the expected number of deaths for a given place and time, can provide a comprehensive account of mortality likely related to the COVID-19 pandemic, including deaths that are both directly and indirectly associated with COVID-19. Since April 2020, CDC’s National Center for Health Statistics (NCHS) has published weekly data on excess deaths associated with the COVID-19 pandemic (1). A previous report identified nearly 300,000 excess deaths during January 26–October 3, 2020, with two thirds directly associated with COVID-19 (2). Using more recent data from the National Vital Statistics System (NVSS), CDC estimated that 545,600–660,200 excess deaths occurred in the United States during January 26, 2020–February 27, 2021.

As of October 15, 216,025 deaths from coronavirus disease 2019 (COVID-19) have been reported in the United States*; however, this number might underestimate the total impact of the pandemic on mortality. Measures of excess deaths have been used to estimate the impact of public health pandemics or disasters, particularly when there are questions about underascertainment of deaths directly attributable to a given event or cause (1-6).(†) Excess deaths are defined as the number of persons who have died from all causes, in excess of the expected number of deaths for a given place and time. This report describes trends and demographic patterns in excess deaths during January 26-October 3, 2020. Expected numbers of deaths were estimated using overdispersed Poisson regression models with spline terms to account for seasonal patterns, using provisional mortality data from CDC's National Vital Statistics System (NVSS) (7). Weekly numbers of deaths by age group and race/ethnicity were assessed to examine the difference between the weekly number of deaths occurring in 2020 and the average number occurring in the same week during 2015-2019 and the percentage change in 2020. Overall, an estimated 299,028 excess deaths have occurred in the United States from late January through October 3, 2020, with two thirds of these attributed to COVID-19. The largest percentage increases were seen among adults aged 25-44 years and among Hispanic or Latino (Hispanic) persons. These results provide information about the degree to which COVID-19 deaths might be underascertained and inform efforts to prevent mortality directly or indirectly associated with the COVID-19 pandemic, such as efforts to minimize disruptions to health care.

BACKGROUND: Pregnancy loss can have physical and psychological consequences for women and their families. Though a previous study described an increase in the risk of self-reported pregnancy loss from 1970 to 2000, more recent examinations from population-based data of US women are lacking. METHODS: We used data from the 1995, 2002, 2006-2010, 2011-2015 National Survey of Family Growth on self-reported pregnancy loss (miscarriage, stillbirth, ectopic pregnancy) among US women (15-44 years) who reported at least one pregnancy conceived during 1990-2011 that did not result in induced termination (n = 20 012 women; n = 42 526 pregnancies). Trends in the risk of self-reported pregnancy loss and early pregnancy loss (<12 weeks) were estimated, separately, by year of pregnancy conception (limited to 1990-2011 to ensure a sufficient sample of pregnancies for each year and maternal age group) using log-Binomial and Poisson models, adjusted for maternal- and pregnancy-related factors. RESULTS: Among all self-reported pregnancies, excluding induced terminations, the risk of pregnancy loss was 19.7% and early pregnancy loss was 13.5% during 1990-2011. Risk of pregnancy loss increased by a relative 2% (rate ratio [RR] 1.02, 95% confidence interval [CI] 1.01, 1.02) per year in unadjusted models and 1% per year (RR 1.01, 95% CI 1.00, 1.02) during 1990-2011, after adjustment for maternal characteristics and pregnancy-related factors. In general, trends were similar for early pregnancy loss. CONCLUSION: From 1990 to 2011, risk of self-reported pregnancy loss increased among US women. Further work is needed to better understand the drivers of this increase in reported pregnancy loss in the US.

Objectives Early pregnancy detection is important for improving pregnancy outcomes as the first trimester is a critical window of fetal development; however, there has been no description of trends in timing of pregnancy awareness among US women. Methods We examined data from the 1995, 2002, 2006-2010 and 2011-2013 National Survey of Family Growth on self-reported timing of pregnancy awareness among women aged 15-44 years who reported at least one pregnancy in the 4 or 5 years prior to interview that did not result in induced abortion or adoption (n = 17, 406). We examined the associations between maternal characteristics and late pregnancy awareness (≥7 weeks' gestation) using adjusted prevalence ratios from logistic regression models. Gestational age at time of pregnancy awareness (continuous) was regressed over year of pregnancy conception (1990-2012) in a linear model. Results Among all pregnancies reported, gestational age at time of pregnancy awareness was 5.5 weeks (standard error = 0.04) and the prevalence of late pregnancy awareness was 23 % (standard error = 1 %). Late pregnancy awareness decreased with maternal age, was more prevalent among non-Hispanic black and Hispanic women compared to non-Hispanic white women, and for unintended pregnancies versus those that were intended (p < 0.01). Mean time of pregnancy awareness did not change linearly over a 23-year time period after adjustment for maternal age at the time of conception (p < 0.16). Conclusions for Practice On average, timing of pregnancy awareness did not change linearly during 1990-2012 among US women and occurs later among certain groups of women who are at higher risk of adverse birth outcomes.

PURPOSE: To evaluate the association between pregnancy loss history and adverse pregnancy outcomes. METHODS: Pregnancy history was captured during a computer-assisted personal interview for 21,277 women surveyed in the National Survey of Family Growth (1995-2013). History of pregnancy loss (<20 weeks) at first parity was categorized in three ways: number of losses, maximum gestational age of loss(es), and recency of last pregnancy loss. We estimated risk ratios for a composite measure of selected adverse pregnancy outcomes (preterm, stillbirth, or low birthweight) at first parity and in any future pregnancy, separately, using predicted margins from adjusted logistic regression models. RESULTS: At first parity, compared with having no loss, having 3+ previous pregnancy losses (adjusted risk ratio (aRR) = 1.66 [95% CI = 1.13, 2.43]), a maximum gestational age of loss(es) at ≥10 weeks (aRR = 1.28 [1.04, 1.56]) or having experienced a loss 24+ months ago (aRR = 1.36 [1.10, 1.68]) were associated with increased risks of adverse pregnancy outcomes. For future pregnancies, only having a history of 3+ previous pregnancy losses at first parity was associated with increased risks (aRR = 1.97 [1.08, 3.60]). CONCLUSION: Number, gestational age, and recency of pregnancy loss at first parity were associated with adverse pregnancy outcomes in U.S. women.

This study aimed to estimate intake of individual polyunsaturated fatty acids (PUFAs), identify major dietary sources of PUFAs and estimate the proportion of individuals consuming fish among US children 12-60 months of age, by age and race and ethnicity. The study employed a cross-sectional design using US National Health and Nutrition Examination Survey data. Representative sample of US population based on selected counties. SUBJECTS: 2496 US children aged 12-60 months. Mean daily intake of n-6 PUFAs and eicosapentaenoic acid (EPA) varied by age, with children 12-24 months of age having lower average intakes (mg or g day(-1) ) than children 49-60 months of age and the lowest n6 : n3 ratio, upon adjustment for energy intake. Docosahexaenoic acid (DHA) intake was low (20 mg day(-1) ) compared to typical infant intake and did not change with age. Compared to non-Hispanic white children, Mexican American children had higher DHA and arachidonic acid (AA) intake. In the previous 30 days, 53.7% of children ever consumed fish. Non-Hispanic black children were more likely than non-Hispanic white children to have consumed fish (64.0% vs. 53.0%). Results indicate low prevalence of fish intake and key n-3 PUFAs, relative to n-6 fatty acids, which suggests room for improvement in the diets of US children. More research is needed to determine how increasing dietary intakes of n-3 PUFAs like DHA could benefit child health.

OBJECTIVES: To describe the contribution of whole fruit, including discrete types of fruit, to total fruit consumption and to investigate differences in consumption by sociodemographic characteristics. METHODS: We analyzed data from 3129 youth aged 2 to 19 years from the National Health and Nutrition Examination Survey, 2011 to 2012. Using the Food Patterns Equivalents Database and the What We Eat in America 150 food groups, we calculated the contribution of whole fruit, 100% fruit juices, mixed fruit dishes, and 12 discrete fruit and fruit juices to total fruit consumption. We examined differences by age, gender, race and Hispanic origin, and poverty status. RESULTS: Nearly 90% of total fruit intake came from whole fruits (53%) and 100% fruit juices (34%) among youth aged 2 to 19 years. Apples, apple juice, citrus juice, and bananas were responsible for almost half of total fruit consumption. Apples accounted for 18.9% of fruit intake. Differences by age were predominately between youth aged 2 to 5 years and 6 to 11 years. For example, apples contributed a larger percentage of total fruit intake among youth 6 to 11 years old (22.4%) than among youth 2 to 5 years old (14.6%), but apple juice contributed a smaller percentage (8.8% vs 16.8%), P < .05. There were differences by race and Hispanic origin in intake of citrus fruits, berries, melons, dried fruit, and citrus juices and other fruit juices. CONCLUSIONS: These findings provide insight into what fruits US youth are consuming and sociodemographic factors that may influence consumption.

BACKGROUND AND OBJECTIVE: Physicians and policy makers are increasingly interested in caffeine intake among children and adolescents in the advent of increasing energy drink sales. However, there have been no recent descriptions of caffeine or energy drink intake in the United States. We aimed to describe trends in caffeine intake over the past decade among US children and adolescents. METHODS: We assessed trends and demographic differences in mean caffeine intake among children and adolescents by using the 24-hour dietary recall data from the 1999-2010 NHANES. In addition, we described the proportion of caffeine consumption attributable to different beverages, including soda, energy drinks, and tea. RESULTS: Approximately 73% of children consumed caffeine on a given day. From 1999 to 2010, there were no significant trends in mean caffeine intake overall; however, caffeine intake decreased among 2- to 11-year-olds (P < .01) and Mexican-American children (P = .003). Soda accounted for the majority of caffeine intake, but this contribution declined from 62% to 38% (P < .001). Coffee accounted for 10% of caffeine intake in 1999-2000 but increased to nearly 24% of intake in 2009-2010 (P < .001). Energy drinks did not exist in 1999-2000 but increased to nearly 6% of caffeine intake in 2009-2010. CONCLUSIONS: Mean caffeine intake has not increased among children and adolescents in recent years. However, coffee and energy drinks represent a greater proportion of caffeine intake as soda intake has declined. These findings provide a baseline for caffeine intake among US children and young adults during a period of increasing energy drink use.

OBJECTIVE: To describe the contribution of mixed dishes to vegetable consumption and to estimate vegetable intake according to specific types of vegetables and other foods among US children and adolescents. DESIGN: The 2003-2008 National Health and Nutrition Examination Survey (NHANES), a nationally representative probability survey conducted in the USA. SETTING: Civilian non-institutionalized US population. SUBJECTS: All children and adolescents aged 2-18 years who met eligibility criteria (n 9169). RESULTS: Approximately 59 % of total vegetable intake came from whole forms of vegetables with 41 % coming from a mixed dish. White potatoes (10.7 (se 0.6) %), fried potatoes (10.2 (se 0.4) %), potato chips (8.6 (se 0.5) %) and other vegetables (9.2 (se 0.5) %) accounted for most vegetables in their whole forms, whereas pasta dishes (9.5 (se 0.4) %), chilli/soups/stews (7.0 (se 0.5) %), pizza/calzones (7.6 (se 0.3) %) and other foods (13.7 (se 0.6) %) accounted for most mixed dishes. Usual mean vegetable intake was 1.02 cup equivalents/d; however, after excluding vegetables from mixed dishes, mean intake fell to 0.54 cup equivalents/d and to 0.32 cup equivalents/d when fried potatoes were further excluded. CONCLUSIONS: Mixed dishes account for nearly half of overall vegetable intake in US children and adolescents. It is critical for future research to examine various components of vegetable intake carefully in order to inform policy and programmatic efforts aimed at improving dietary intake among children and adolescents.

Adequate folate and iron intake during pregnancy is critical for maternal and fetal health. No previous studies to our knowledge have reported dietary supplement use and folate status among pregnant women sampled in NHANES, a nationally representative, cross-sectional survey. We analyzed data on 1296 pregnant women who participated in NHANES from 1999 to 2006 to characterize overall supplement use, iron and folic acid use, and RBC folate status. The majority of pregnant women (77%) reported use of a supplement in the previous 30 d, most frequently a multivitamin/-mineral containing folic acid (mean 817 mug/d) and iron (48 mg/d). Approximately 55-60% of women in their first trimester reported taking a folic acid- or iron-containing supplement compared with 76-78% in their second trimester and 89% in their third trimester. RBC folate was lowest in the first trimester and differed by supplement use across all trimesters. Median RBC folate was 1628 nmol/L among users and 1041 nmol/L among nonusers. Among all pregnant women, median RBC folate increased with trimester (1256 nmol/L in the first, 1527 nmol/L in the second, and 1773 nmol/L in the third). Given the role of folic acid in the prevention of neural tube defects, it is notable that supplement use and median RBC folate was lowest in the first trimester of pregnancy, with 55% of women taking a supplement containing folic acid. Future research is needed to determine the reasons for low compliance with supplement recommendations, particularly folic acid, in early pregnancy.

Among children with food allergy, we aim to describe differences in allergy severity by sociodemographic characteristics and potential differences in healthcare characteristics according to food allergy severity. Using the 2007 National Survey of Children's Health, we identified children with food allergies based on parental report (n = 4,657). Food allergic children were classified by the severity of their food allergy, as either mild (n = 2,333) or moderate/severe (n = 2,285). Using logistic regression, we estimated the odds of having moderate/severe versus mild food allergy by sociodemographic characteristics and the odds of having selected healthcare characteristics by food allergy severity. Among children with food allergy, those who were older (ages 6 through 17 years) and those who had siblings were more likely to have moderate/severe allergy compared to their younger and only-child counterparts. There were no significant differences in severity by other sociodemographic characteristics. Children with a moderate/severe food allergy were more likely to report use of an Individual Education Plan (OR = 1.88 [1.31, 2.70]) and to have seen a specialist than those with mild food allergy. Among younger children with food allergy, those with moderate/severe food allergy were more likely to require more services than is usual compared with those with mild allergy. Associations between allergy severity and health care-related variables did not differ significantly by race/ethnicity, income level, or maternal education. We report few differences in allergy severity by sociodemographic characteristics of food allergic children. In addition, we found that associations between allergy severity and use of health related services did not differ significantly by race/ethnicity or poverty status among children with food allergy. Given the importance of food allergy as an emerging public health issue, further research to confirm these findings would be useful.

The Journal recently published an article suggesting a causal association between electronic fetal monitoring (EFM) and decreased infant mortality.1 We have strong reservations regarding the ability of that paper to offer guidance regarding the effectiveness of EFM because it is inappropriate to use vital statistics data to make the leap from statistical association to causation. | Several years ago, two of us contributed to a set of American Journal of Obstetrics and Gynecology commentaries discussing the reasonable use of secondary vital statistics data.2, 3 Those commentaries raised important issues and limitations that should be considered, acknowledged, and addressed.

There is increasing evidence that in utero effects of excessive gestational weight gain may result in increased weight in children; however, studies have not controlled for shared genetic or environmental factors between mothers and children. Using 2,758 family groups from the Collaborative Perinatal Project, the authors examined the association of maternal prepregnancy body mass index (BMI) and gestational weight gain on child BMI at age 4 years using both conventional generalized estimating equations and fixed-effects models that account for shared familial factors. With generalized estimating equations, prepregnancy BMI and gestational weight gain had similar associations with the child BMI z score (beta = 0.09 units, 95% confidence interval (CI): 0.08, 0.11; and beta = 0.07 units, 95% CI: 0.04, 0.11, respectively. However, fixed effects resulted in null associations for both prepregnancy BMI (beta = 0.03 units, 95% CI: -0.01, 0.07) and gestational weight gain (beta = 0.03 units, 95% CI: -0.02, 0.08) with child BMI z score at age 4 years. The positive association between gestational weight gain and child BMI at age 4 years may be explained by shared family characteristics (e.g., genetic, behavioral, and environmental factors) rather than in utero programming. Future studies should continue to evaluate the relative roles of important familial and environmental factors that may influence BMI and obesity in children.

OBJECTIVES: The goals were to estimate the prevalence of food allergy and to describe trends in food allergy prevalence and health care use among US children. METHODS: A cross-sectional survey of data on food allergy among children <18 years of age, as reported in the 1997-2007 National Health Interview Survey, 2005-2006 National Health and Nutrition Examination Survey, 1993-2006 National Hospital Ambulatory Medical Care Survey and National Ambulatory Medical Care Survey, and 1998-2006 National Hospital Discharge Survey, was performed. Reported food allergies, serum immunoglobulin E antibody levels for specific foods, ambulatory care visits, and hospitalizations were assessed. RESULTS: In 2007, 3.9% of US children <18 years of age had reported food allergy. The prevalence of reported food allergy increased 18% (z = 3.4; P < .01) from 1997 through 2007. In 2005-2006, serum immunoglobulin E antibodies to peanut were detectable for an estimated 9% of US children. Ambulatory care visits tripled between 1993 and 2006 (P < .01). From 2003 through 2006, an estimated average of 317000 food allergy-related, ambulatory care visits per year (95% confidence interval: 195000-438000 visits per year) to emergency and outpatient departments and physician's offices were reported. Hospitalizations with any recorded diagnoses related to food allergy also increased between 1998-2000 and 2004-2006, from an average of 2600 discharges per year to 9500 discharges per year (z = 3.4; P < .01), possibly because of increased use of food allergy V codes. CONCLUSION: Several national health surveys indicate that food allergy prevalence and/or awareness has increased among US children in recent years.

BACKGROUND: The role of intergenerational influences on age at menarche has not been explored far beyond the association between mothers' and daughters' menarcheal ages. Small size at birth and childhood obesity have been associated with younger age at menarche, but the influence of maternal overweight or obesity on daughters' age at menarche has not been thoroughly examined. METHODS: In a follow-up study of the prospective Collaborative Perinatal Project, grown daughters were asked in 1987-1991 for their age at menarche. Data from the original Collaborative Perinatal Project (1959-1966) included their mothers' height and prepregnancy weight. In the follow-up study, 597 of 627 daughters had complete menarche and maternal data available and were included in the present analysis. We used polytomous logistic regression to examine the association between maternal overweight (body mass index [BMI] = 25-29.9 km/m) or obesity (BMI >or= 30) and daughter's age at menarche (<or=12, 12, 13, 14+ years). RESULTS: Compared with those whose mothers had a BMI less than 25, daughters of obese mothers experienced younger age at menarche (OR for menarche at <or=12 years = 3.1 [1.1-9.2]). This association remained after adjusting for maternal age at menarche, maternal parity, socioeconomic status, race, and study site (OR = 3.3 [1.1-10.0]). Effect estimates for maternal overweight were close to the null. There was limited evidence of mediation by small for gestational age or BMI at age 7. CONCLUSIONS: Maternal obesity is associated with younger menarcheal age among daughters in this study, possibly via unmeasured shared factors.

BACKGROUND: The sex ratio in the USA has declined over recent decades, resulting in fewer male births. Concurrent changes in the childbearing population may have influenced the sex ratio, including increases in multiple births, improvements in perinatal survival and increased Hispanic births. METHODS: Data from the US natality files (1981-2006) were analyzed to determine the impact of changes in birth characteristics on male birth proportion. Male birth proportion was calculated as the number of male births divided by the total number. In separate analyses, trends in male birth proportion from 1981 to 2006 were adjusted for plurality (singleton, multiple), gestational age (<28, 28-32, 33-36, ≥37 weeks) and, from 1989, maternal Hispanic ethnicity. Separate analyses were conducted for white and black births. Log binomial regression was performed to estimate crude and adjusted trends with year as independent variable. RESULTS: Trends in male birth proportion differed significantly according to plurality among white (P < 0.01), but not black births. Adjustment for gestational age tempered the trends among white singletons (P < 0.0001) and multiples (P < 0.05) but had no effect on trends in black male birth proportion. Adjustment for Hispanic ethnicity had no impact on trends in black male birth proportion and any effect on white births was negated by changes in gestational age trends. CONCLUSIONS: Lack of consistent influences on, or patterns of change in, the proportion of male births between different subpopulations of births suggests that a single mechanism is unlikely to explain the oft-referenced decrease in the overall US sex ratio.