There is evidence for long-lasting effects of birth characteristics on cognitive ability in childhood and adulthood. Further, low cognitive ability throughout the lifetime has been linked to age-related cognitive decline and dementia risk. However, little is known about the effects of birth characteristics on cognitive dysfunction late in life. Here we explore potential associations between birth characteristics , adjusted and not adjusted for gestational age, and cognitive impairment and dementia late in life.p < 0.001) compared to infants with normal growth and head size, even after controlling for familial factors, childhood socioeconomic status, and education in adulthood. In line with this, each additional 100 g birth weight and each additional millimeter head circumference significantly reduced the risk for dementia  and cognitive impairment , respectively. Within-pair analyses of identical twins, though hampered by small sample size, suggested that the observed associations between birth characteristics and dementia are likely not due to underlying shared genetic or environmental etiology. A limitation of the present study is that registry-based dementia diagnoses likely miss some of the true dementia cases in the population. Further, a more precise measure of cognitive reserve early in life as well as a date of onset for the cognitive impairment measure in the subsample would have been favorable.Data from twins in the Swedish Twin Registry born 1926–1960 were merged with information from the Swedish birth, patient, and cause of death registries, resulting in a sample of 35,191 individuals. A subsample of 4,000 twins aged 65 years and older also participated in a telephone cognitive screening in 1998–2002. Associations of birth characteristics with registry-based dementia diagnoses and on telephone-assessed cognitive impairment were investigated in the full sample and subsample, respectively. The full sample contained 907 (2.6%) individuals with a dementia diagnosis , 803 (2.4%) individuals born small for gestational age, and 929 (2.8%) individuals born with a small head for gestational age. The subsample contained 569 (14.2%) individuals with cognitive impairment. Low birth weight for gestational age and being born with a small head for gestational age were significant risk factors for cognitive dysfunction late in life, with an up to 2-fold risk increase  have a significantly higher risk of age-related cognitive dysfunction compared to those with normal growth, highlighting the importance of closely monitoring the cognitive development of such infants and evaluating the potential of early life interventions targeted at enhancing cognitive reserve. Using data from the Swedish Birth Registry, Miriam Mosing and colleagues investigate associations between healthy prenatal growth and cognitive dysfunction in late-life. There is evidence for associations between birth characteristics and cognitive ability in childhood and adulthood.Low cognitive ability throughout the lifetime has been linked to age-related cognitive decline and dementia risk.Little is known about associations between birth characteristics and cognitive dysfunction late in life, such as dementia and cognitive impairment.We conducted a registry-based cohort study of 35,191 individuals linking data from the Swedish birth registry to the Swedish patient, cause of death, and twin registries to explore potential associations between birth characteristics and age-related cognitive dysfunction as measured by all-cause dementia and cognitive impairment.Low birth weight and small head circumference pose a significant risk for late life cognitive dysfunction.The findings highlight the importance of healthy prenatal growth and closely monitoring the cognitive development of small infants and those with signs of poor fetal growth.Potential interventions to increase cognitive reserve could be considered for infants with low birth weight and small head circumference. Associations between cognitive function and anthropometric measures in childhood and adulthood, such as body height and head circumference, are well documented. Several studies have reported positive associations of cognitive ability with body length in childhood  and in aFurther, it has repeatedly been shown that adults with higher cognitive ability, higher education, and higher occupational achievements seem to be somewhat protected from developing age-related cognitive impairments such as dementia –16. To aFinally, birth characteristics, such as preterm birth, low birth weight, short birth length, small head circumference, and indicators of reduced fetal growth , are independently associated with lower academic achievement, cognitive function, and intellectual performance in childhood ,20–22 anTogether, the well-documented relationships between size at birth, gestational age, and cognition, as well as adult cognition and dementia, and the associations between adult anthropometric measures and dementia imply potentially adverse effects of reduced fetal growth for age-related cognitive decline and dementia. At this time, when population aging is one of the most important demographic phenomena in the world , and givA link between birth characteristics and age-related cognitive dysfunction could in part explain the reported associations between adult anthropometric measurements, cognitive ability, and age-related cognitive dysfunction. Understanding these relationships is essential for the development of prevention strategies and to identify individuals who may benefit from early intervention. The present study explores potential associations between birth characteristics—i.e., weight, length, head circumference, and gestational age, as absolute measures as well as relative to gestational age—and all-cause dementia, using a large Swedish population-based study of more than 35,000 individuals. In addition, as registry-based dementia diagnoses may underestimate true numbers of cases of dementia in the population , associaThe data were based on all twins registered with the Swedish Twin Registry (STR) ,33 born We excluded 55 individuals who had passed away before data from the NPR were available and 21 individuals with early onset dementia (onset before the age of 55 years) and 33 individuals because of Down syndrome, mental retardation, or cerebral palsy. An additional 57 individuals with impossible gestational age–birth weight combinations  were excN = 20,269) participating in the Screening Across the Lifespan Twin (SALT) study, conducted between 1998 and 2002 with all then living members of the STR [Cognitive screening data from a telephone-based interview were available from all twins who were aged 65 years and older ( the STR . A subsa the STR .http://ki.se/en/research/swedish-twin-registry-for-researchers), the NPR , and the CDR .This study was approved by the Regional Ethical Review Board in Stockholm . Because this study was strictly register-based, individual informed consent was not deemed necessary . DetailsBirth characteristics for the twins were derived from official birth records. Birth information was based on a nationwide collection (starting in 1926) of information from original birth records recorded by midwives and/or doctors at the time of birth. Birth record data were linked to collected data from the STR using the unique personal identification number assigned to all Swedish citizens. To minimize potential misclassification, only twins with known birth order were included. For twins born between 1926 and 1958, birth order was ascertained if the twins’ names could be found in the medical record or if they provided information on birth order in a later telephone interview. For twins born between 1959 and 1960, correct birth order was ascertained using an algorithm based on information from the medical record  in combination with information on birth weight and birth order from later data collections in the STR. For further details on the matching procedure see Hogberg and colleagues .BW was explored as continuous raw weight scores  as well as dichotomized for low birth weight (LBW), coded as low  versus normal . Further, 2 weight variables were derived adjusted for gestational age (GA) in days: (1) continuous birth weight adjusted for gestational age (BWGA) and sex, which was derived by regressing GA on BW and then standardizing the residuals by sex, and (2) small for gestational age (SGA), a statistical cutoff for the dynamic concept of poor fetal growth calculated here using the common Swedish definition of weighing less than 2 SD below the mean for a given GA and sex. Note that SGA is not synonymous with fetal growth restriction, where the fetus fails to reach its growth potential. Infants classified as SGA could be in this category for the following 3 reasons: (1) a non-malformed infant not having reached its growth potential , (2) a malformed infant for whom the fetal growth pattern is linked to malformations, and (3) a genetically small but otherwise normal infant. As the BW data nicely followed a normal distribution, and so as not to lose valuable information, extreme outliers more than 4 SD above/below the mean  were Winsorized, i.e., set to 4 SD above or below the mean .The potential predictive value of HC was explored as continuous raw variable (in millimeters), as well as adjusted for GA (in days) and sex (see above) as a continuous variable  and dichotomized for small for a given gestational age and sex (less than 2 SD below the mean) . As above, only 9 extreme outliers (more than 6 SD above/below the mean) were Winsorized .BL was similarly coded as a continuous raw variable (in centimeters), as well as adjusted for GA (in days) and sex , and dichotomized for short for a given GA and sex (less than 2 SD below the mean) . Six outliers (more than 5 SD below the mean) were Winsorized .GA was coded as the number of weeks from the first day of the last menstrual period until birth. In addition, a dichotomous variable called preterm was derived, and individuals born before 37 gestational weeks were considered preterm.Register-based dementia diagnoses (coded as a dichotomous variable) as well as date of first diagnosis were derived from the NPR and CDR, including all primary and contributory diagnoses in the NPR and all underlying and contributory causes of death in the CDR. The various ICD codes used to detect the different types of dementia  are shown in Computer-assisted telephone cognitive screening was done in 2 stages. First, participants were interviewed with an instrument known as TELE ,40. The In addition to sex and age/year of birth  of the participants, a number of other factors that could potentially also influence birth characteristics and age-related cognitive dysfunction were included as covariates where applicable. For descriptive information on covariates see Parity (the number of children previously born to the same mother) and age of mother at birth were included as covariates where applicable.Birth SES is a well-known predictor of child cognitive development, with lower SES being associated with lower IQ ,46, and Education level of the participants was also included as it has been shown to be a strong marker of cognitive reserve . EducatiNs in those models corrected for increasing numbers of covariates, the multiple imputation (MI) procedure in StataIC 14 [mi impute chained), a MI method using a sequence of univariate imputations with fully conditional specification of prediction equations [As some of the variables (see below) had missing values, resulting in reduced ataIC 14  was usedataIC 14 . That waataIC 14  was usedquations . The augThough there was no specific study protocol for the present study, all analyses were carefully planned with a statistician and decided on before the analyses were conducted. The only data-driven change to the analyses was the MI process, conducted to account for missingness in some of the covariates. While originally only 10 and then 40 datasets were imputed, this was increased to 58 and 100 for the full sample and subsample, respectively, following recommendations by von Hippel  and GrahSurvival analyses were conducted to explore the association between birth characteristics  and dementia. The number of months of survival (age) from date of birth to date of dementia diagnosis or date of censoring  was used as the time scale . Cox proportional hazards regression analysis was used to calculate hazard ratios (HRs). HRs represent the effect of a 1-unit change in the predictor  on the base dementia risk (raw hazard of dementia) during the follow-up period. Study participants were followed from 1977 (when data from the NPR were available). To correct for correlation within twin pairs, the robust standard error estimator for clustered observations was used. Four models were fitted for each of the birth characteristics: (1) without any covariates; (2) corrected for sex, YOB, parity, and age of mother; (3) corrected for birth SES in addition to the covariates included in model 2; and (4) corrected for education level in addition to all covariates included in model 3. The assumption of proportional hazards was tested for each model using Schoenfeld residuals . No evidTo obtain odds ratios (ORs), logistic regressions were conducted with the dichotomized cognitive impairment score as outcome variable and BW , HC , BL , and GA (continuous in weeks and preterm) as the predictors. Again, robust estimators of standard errors for clustered data were used. Initially, a baseline model was fitted for each of the birth characteristics and the outcome variable separately without correcting for any of the covariates. The analyses were then repeated including the various covariates, i.e., age (at cognitive screening), sex, parity, and age of mother, birth SES, and education level (see above description of the 4 models).To further explore associations with regard to familial  factors shared within the twin pairs, within-pair (co-twin control) analyses in identical pairs were conducted. Identical  twins share all their segregating genes as well as their family environment. If adverse birth effects cause a risk for cognitive dysfunction, we would expect that a twin with more unfavorable birth characteristics would have a higher risk for cognitive dysfunction than his or her co-twin with more favorable birth characteristics . Note that only complete identical twin pairs discordant for exposure  and outcome  contributed to the within-pair analyses. Conditional (Cox) regression models with family identification as the stratum variable were conducted for dementia, thereby fixing individual baseline hazard within pairs while allowing it to vary between pairs . In a maOne of reviewer’s comments led us to investigate further whether birth order within pairs may play a role in late life cognitive dysfunction. The second-born twin is frequently the smaller one and also has a higher risk for asphyxia, which may result in lower cognitive ability and could therefore mediate potential associations between birth characteristics and cognitive dysfunction late in life. We therefore tested whether birth order (corrected for sex) had a main effect on the 2 cognitive outcomes and then repeated the imputation process and all main analyses, including the co-twin control analyses, adjusting for birth order.p > 0.5) for dementia and cognitive impairment, suggesting no association between birth order and cognitive dysfunction.The final dataset for the birth characteristics–dementia analyses contained 35,191 twin individuals aged 55 to 89 years  at follow-up, with information on at least 1 of the birth characteristics and dementia diagnosis . BW rangp < 0.05) and later birth decade  somewhat increasing the risk for dementia, i.e., those born between 1936 and 1945 showed a somewhat increased risk (non-significant), while those born between 1946 and 1955 had an about 40% higher risk compared to those born 1926 to 1935 (p < 0.05). There was little evidence in support of an association between birth SES or education and dementia (even when included without any of the birth-related variables). Note that overall results based on the full unimputed sample as well as the reduced sample  were very similar in terms of both effect sizes and significance levels, although the latter varied slightly between analyses depending on the sample size  and HCGA, and dementia, infants with SHCGA had an about 65% higher risk for dementia, independent of the covariates included. Though the estimates were in the expected direction for all 3 BL variables, with an up to 40% increased risk for dementia for those born with SBLGA, none of the estimates were significant. Similarly, there was little evidence in support of an association between GA or preterm and dementia. Of the covariates, parity and birth decade were the only significant risk factors for dementia, with increased parity (size see  Tables. size see .p < 0.01) and parity number (p < 0.01), as well as a 35% lower odds for those with higher education (p < 0.001). Again, results based on the full unimputed sample as well as the reduced sample  were very similar in terms of both effect sizes and significance levels, although the latter varied slightly between analyses depending on the sample size . Of the covariates, age, parity, and education were significant, with 7% higher odds of being diagnosed with cognitive impairment with each additional year of age  to explore whether the significant associations observed in the above analyses were confounded by underlying shared liability  and dementia and cognitive impairment later in life. Overall, lower BW and smaller HC showed associations in the expected direction, resulting in increased risks for age-related cognitive dysfunction, i.e., dementia or cognitive impairment. For dementia, lower BW (independent of GA) was a significant risk factor, with a 2% higher risk with each 100 g lower BW and about 22% higher risk if weighing less than 2.5 kg at birth. Similar associations were observed when BW was adjusted for GA, with a 9% risk reduction for dementia for each additional standard deviation BWGA and a 17% risk increase for individuals born SGA, though the latter was not significant. HC was a significant risk factor only when the head was SGA (65% risk increase), suggesting harmful effects of reduced fetal growth for dementia. Similarly, detrimental effects were observed for cognitive impairment, with individuals born SGA and with SHCGA showing approximately 73% and 124% higher odds, respectively. In addition, larger HC was, independent of GA as well as adjusted for GA, a significant protective factor for cognitive impairment (but not for dementia), with each additional millimeter resulting in a 1% lower odds for cognitive impairment and each additional SD HCGA resulting in a 15% lower odds. BL and GA (independent of other birth characteristics) showed no significant association with late life cognitive dysfunction. Overall, these findings suggest negative long-term effects of small birth size on risk of cognitive dysfunction late in life, with additional protective effects of each extra 100 g of BW for dementia and each additional millimeter of HC for cognitive impairment (independent of GA).Past research has shown birth characteristics adjusted and unadjusted for GA to be independently associated with academic achievement, cognitive function, and intellectual performance early in life ,20–25, tContrary to prior literature and the findings of another study using a partly overlapping sample in the STR ,60, educOf the other covariates, only higher parity and later birth decade were significant for dementia, and increased age and parity for cognitive impairment. The finding that higher parity increases the risk for cognitive dysfunction later in life is in line with previous findings showing an increased risk for low intellectual performance with increased parity . FurtherEffect sizes of the within-pair analyses, which are adjusted for shared influences, were of similar magnitude (though not significant) to those from between-pair analyses, suggesting that the increased risk for cognitive dysfunction for individuals who were of smaller birth size is likely not explained by underlying shared etiology, such as genetic factors or shared environmental influences . These findings, though hampered by a small sample size and low power, could suggest that the smaller twin was exposed to a less optimal intrauterine environment, which manifested itself in the less ideal birth characteristics as well as a worse trajectory throughout the lifetime in terms of cognitive development and cognitive aging. These results are in line with past reports of effects of poor fetal growth (as indicated by adverse birth characteristics) on cognitive development in early childhood based on co-twin control analyses . HoweverIn light of the past literature, our findings show that small birth size and low fetal growth may increase the risks of age-related cognitive dysfunction late in life even when controlling for familial factors and childhood SES as well as education in adulthood. Further, each additional 100 g BW and additional millimeter HC significantly reduced the risk of dementia and cognitive impairment, respectively, independent of as well as adjusted for GA. Although our findings suggest that the frequently reported protective value of cognitive reserve (as measured by cognitive ability) on cognitive dysfunction may at least to some degree be explained by factors established as early as during fetal development, the increased risk of age-related cognitive dysfunction associated with small birth size and the indicators of fetal growth measured here was also evident in models corrected for childhood SES and education—predictor and proxy of cognitive ability, respectively. It is known that a multitude of factors influence age-related cognitive decline (as well as cognitive reserve), including genetic and neurological factors, immune response, and various environmental factors such as nutrition, physical activity, family environment, physical and mental trauma, and occupation, which are likely to interact with each other throughout the lifetime. It is probable that exposure to a less optimal intrauterine environment will act negatively on several of the factors named above, including general health as well as brain and cognitive reserve, although it is yet to be determined whether the associations of small birth size and low fetal growth with cognitive decline are mediated by any of those factors. Our findings highlight the importance of preventing prenatal growth restriction and closely monitoring the cognitive development of growth-restricted infants. It has been shown that early life catchup growth may be beneficial for the development of those born with adverse birth characteristics , and groAs touched upon above, there are some limitations to the use of registry-based data. Among these is the lack of sensitivity of dementia diagnoses, as well as the possibility of mistakes in the birth records . Given this, it is even more remarkable that such strong associations between birth characteristics and dementia were observed in the registry data and that the findings could be replicated for cognitive impairment, which was not based on medical records but on a cross-sectional cognitive screening study. As a result, though, the cognitive impairment data had no time-to-event information , which would have been desirable, and had to be modeled as ORs rather than HRs. Another potential caveat of registry-based dementia diagnosis is the potential for misdiagnosis, which has been shown to be especially high in those below the age of 65 years . HoweverIn addition, the dating (exact GA) of pregnancies would likely have been less accurate at the time the birth data were recorded compared to today, which could potentially explain the non-significant associations observed for GA and some GA-corrected variables. Further, perinatal complications, such as asphyxia, have been associated with developmental deficits in both premature and SGA infants , which ip-values. An assumption of MI methods is that the data are missing at random. The greatest extent of missingness was found for birth SES. Should the data not be missing at random, it is likely that those missing would be at the lower end of the distribution, potentially resulting in residual confounding from birth SES of the effect sizes. Nevertheless, as mentioned above, the distributions of the birth characteristics and covariates in the present sample were in line with those previously reported for twin samples [Not unexpectedly given the age of the records, there was information missing in the available birth data. In order to obtain less biased and more precise estimations of the parameters of interest, MI was conducted, which allowed us to include those individuals who would have been omitted in a complete case analysis. However, the overall rate of missingness was only 6.2% in the full sample and 9.7% in the subsample, which is generally regarded as inconsequential . We also samples . Finally samples ), raisin samples ,75, sugg samples –78). As In conclusion, in this study, we found that smaller birth size  is associated with a significant risk for late life cognitive dysfunction, largely independent from education level and childhood SES—factors related to cognitive ability. As such, our findings highlight the importance of preventing prenatal growth restriction, closely monitoring the cognitive development of growth-restricted infants, and exploring the potential of reserve-enhancing interventions to possibly protect individuals at higher risk for late life cognitive dysfunction.S1 STROBE Checklist(PDF)Click here for additional data file.S1 Table(DOCX)Click here for additional data file.S2 Table(DOCX)Click here for additional data file.S3 Table(DOCX)Click here for additional data file.S4 Table(DOCX)Click here for additional data file.S5 Table(DOCX)Click here for additional data file.S6 Table(DOCX)Click here for additional data file.S7 Table(DOCX)Click here for additional data file.S8 Table(DOCX)Click here for additional data file.S9 Table(DOCX)Click here for additional data file.


Previous literature suggests a U-shaped relation between hemoglobin concentration and adverse birth outcomes. There is less evidence on associations between iron status and birth outcomes.Our objective was to determine the associations of maternal hemoglobin concentration and iron status with birth outcomes.n = 1137) and Malawi (n = 1243). Hemoglobin concentration and 2 markers of iron status [zinc protoporphyrin and soluble transferrin receptor (sTfR)] were measured at ≤20 weeks and 36 weeks of gestation. We used linear and Poisson regression models and birth outcomes included preterm birth (PTB), newborn stunting, low birth weight (LBW), and small-for-gestational-age.We conducted a secondary data analysis of data from 2 cohorts of pregnant women receiving iron-containing nutritional supplements (20–60 mg ferrous sulfate) in Ghana  at enrollment was 9% in Ghana and 20% in Malawi. In early pregnancy, iron deficiency was associated with PTB  and stunting  in Malawi but not Ghana, and was not associated with LBW in either country; replete iron status (sTfR <10th percentile) was associated with stunting  in Ghana, but not PTB or LBW, and was not associated with any birth outcomes in Malawi. In late pregnancy, iron deficiency was not related to birth outcomes in either country and iron-replete status was associated with higher risk of LBW  and stunting  in Ghana, but was not associated with birth outcomes in Malawi.clinicaltrials.gov as NCT00970866 (Ghana) and NCT01239693 .The associations of low or replete iron status with birth outcomes are population specific. Research to replicate and extend these findings would be beneficial. These trials were registered at  The WHO currently recommends daily supplementation with 30–60 mg/d elemental iron (+400 μg folic acid) throughout pregnancy, and in settings where anemia in pregnant women is a severe public health problem , the daily dose of 60 mg is recommended over a lower dose . AlthougNumerous studies have reported a U-shaped relation between maternal hemoglobin (Hb) concentrations during pregnancy and adverse birth outcomes, with both low and high Hb concentrations associated with higher risks of low birth weight (LBW) and preterm birth (PTB) , althougA recent review noted that in studies that have examined maternal iron status and birth outcomes, serum ferritin concentration is the most widely used biomarker of iron status . Serum fLittle research has been conducted regarding maternal iron status and birth outcomes in sub-Saharan Africa , despitewww.ilins.org). The primary objective of these trials was to determine the efficacy of small-quantity lipid-based nutrient supplements (SQ-LNSs) for preventing malnutrition in pregnant and lactating women and their infants. Details of the study methods have been reported elsewhere (1) a 60-mg Fe (ferrous sulfate) +400-µg folic acid capsule; 2) a multiple micronutrient capsule (18 micronutrients); or 3) a sachet of SQ-LNS . Both the multiple micronutrient capsule and SQ-LNS contained 20 mg Fe (ferrous sulfate) per daily supplement. Differences in inclusion and exclusion criteria included maternal age  and HIV status . Women received intermittent preventive malaria treatment in accordance with standard antenatal care recommendations in Ghana and Malawi. At both enrollment and 36 weeks of gestation, trained phlebotomists collected blood samples by venipuncture with heparinized blood collection tubes. Blood samples were typically collected in the morning, with the majority of samples collected between 0800 and 1200. Previous studies suggest that time of day does not affect the reliability of markers of iron status  and Ghana (NCT00970866), both malaria-endemic countries, as part of the International Lipid-Based Nutrient Supplements Project (lsewhere . Brieflyn status . In Ghan1) at least one measurement of Hb concentration or iron status, at either enrollment or 36 weeks of gestation, and 2) any newborn anthropometric measurement or an estimated pregnancy duration. We have previously reported that supplementation with SQ-LNSs or multiple micronutrients resulted in lower Hb and iron status at 36 weeks of gestation than for the iron and folic acid trial arm . Plasma samples were stored at −20°C and shipped to the USDA Agricultural Research Service Western Human Nutrition Research Center, Davis, CA where sTfR (mg/L), α-1-acid glycoprotein [AGP (g/L)], and C-reactive protein [CRP (mg/L)] concentrations were determined using a Cobas Integra 400 plus Automatic Analyzer (Roche Diagnostic Corp.). The inter- and intra-assay CVs for these assays were: 1) sTfR intra-assay: <2.2%, interassay: <1.1%; 2) AGP intra-assay: <2.9%, interassay: <2.0%; 3) CRP: intra-assay: <3.5%, interassay: <1.3%.Laboratory technicians assayed Hb (HemoCue) and malaria parasitemia with rapid tests  in peripheral blood and centrifuged the remaining blood samples at 1252 × We tested normality using the Shapiro–Wilk test. Hb was determined to have a normal distribution and ZPP and sTfR concentrations were log transformed. We also analyzed Hb, ZPP, and sTfR using categorical variables. We defined anemia using a cutoff value of Hb <100 g/L, based on research suggesting that this cutoff value is more accurate when defining anemia in pregnant women of African descent , 25, 26.z score (LAZ), and head circumference z score (HCZ)]. Poisson regression models were used to estimate RR for dichotomous birth outcomes, including PTB (<37 weeks of gestation), LBW (<2.5 kg), small-for-gestational-age (SGA) [birth weight <10th percentile by gestational age and sex using the INTERGROWTH-21st standard (P < 0.1) associated with the outcome in bivariate analyses. Based on previous literature, variables identified a priori as potential confounding factors were gestational age at enrollment, parity (nulliparous compared with parous), maternal age, education level, household food insecurity, household asset index, AGP at the time the blood sample was drawn, CRP at the time the blood sample was drawn, infant sex, baseline maternal BMI (in kg/m2), a positive rapid test for malaria at enrollment, and HIV status . Household food insecurity was measured using a scale ranging from 0 (no food insecurity) to 27 (every food insecurity condition occurs often)  . We creas often) . SpecifiP < 0.1, and stratified analyses were performed for significant interactions. All analyses were performed using SAS version 9.4 (SAS Institute).Interactions with maternal age, parity, and supplement group were examined and determined to be significant at n = 22, Malawi: n = 12), miscarriage , stillbirth , and loss-to-follow-up  (P < 0.001); have a lower proxy SES ; and be less often nulliparous , HIV positive , or anemic  than excluded participants, but they had similar BMI, years of education, and malaria rates at enrollment to those excluded. In Ghana, excluded study participants did not differ from included participants in any of the aforementioned listed characteristics. Study participant characteristics are provided in Of the 1320 and 1391 women enrolled in the trials in Ghana and Malawi, respectively, 1137 and 1243 were included in this set of analyses, respectively. Reasons for exclusion included twin pregnancy , iron deficiency (sTfR >6.0 mg/L), and IDA and mean concentrations of Hb, sTfR, and ZPP at enrollment and 36 weeks of gestation are presented in At enrollment, Hb concentration in adjusted linear regression analyses was not associated with any birth outcomes in Ghana but was associated with a longer pregnancy duration, higher birth weight, and higher newborn LAZ in Malawi (P = 0.046) and iron deficiency showed a similar trend (P = 0.07). High Hb (>130 g/L) was not associated with adverse birth outcomes in either Ghana or Malawi  and poorer birth outcomes in Ghana. Lower sTfR concentration can be a sign of impaired RBC production (which is linked to inadequate PVE)   and possOne potential biological mechanism that could explain why higher iron status may be detrimental to the fetus is non–transferrin bound iron. Iron is typically carefully chaperoned around the body, predominately by transferrin. Unbound iron can result when the rate of iron influx into plasma exceeds the rate of iron acquisition by transferrin . It is tAssociations between iron status and birth outcomes may have differed by country owing to differences between Ghanaian and Malawian women in iron status at enrollment. Ghanaian women had higher iron status  than Malawian women at enrollment, and a lower proportion of Ghanaian women were identified as iron deficient . All women in these 2 cohorts received iron supplementation, with approximately two-thirds receiving 20 mg and one-third receiving 60 mg Fe/d. Thus there was a larger number of iron-replete women in Ghana than in Malawi who received an amount of iron supplementation that may have been unnecessary, and plausibly had negative effects on the fetus. We previously demonstrated that birth weight was higher among infants born to women receiving SQ-LNSs (which contained 20 mg Fe) than among infants of women receiving iron and folic acid capsules (which contained 60 mg Fe), even though the iron and folic acid group had higher mean Hb concentration, higher iron status, and a lower prevalence of anemia at 36 weeks of gestation , 22. It Strengths of this study include the use of ultrasound to estimate gestational age at enrollment; the availability of a large number of covariates to test and control for confounding, including 2 markers of inflammation; and analysis of 2 cohorts from different regions in Africa where similar study methods were used. Including markers of inflammation is important, because inflammation can lead to elevated sTfR concentrations . In addiWe are limited in the interpretation of our results, because it is unclear whether low sTfR can be used as an indicator of iron-replete status. There is evidence that sTfR is lower during impaired erythropoiesis , which cIn conclusion, this research provides evidence that the associations between low or replete maternal iron status and birth outcomes are population specific. Future research to replicate and extend these findings would be beneficial.nxy278_Supplemental_FilesClick here for additional data file.


Common exercises such as the barbell back squat (BBS) and barbell hip thrust (BHT) are perceived to provide a training stimulus to the lumbar extensors. However, to date there have been no empirical studies considering changes in lumbar extension strength as a result of BBS or BHT resistance training (RT) interventions.To consider the effects of BBS and BHT RT programmes upon isolated lumbar extension (ILEX) strength.n = 14; 22.07 ± 0.62 years; 179.31 ± 6.96 cm; 79.77 ± 13.81 kg) were randomised in to either BBS (n = 7) or BHT (n = 7) groups and performed two training sessions per week during a 4-week mesocycle using 80% of their 1RM. All subjects were tested pre- and post-intervention for BBS and BHT 1RM as well as isometric ILEX strength.Trained male subjects . And the BHT group improved their BHT 1RM to a greater degree than the BBS group . Neither BBS nor BHT groups significantly improved their isometric ILEX strength.Analyses revealed that both BBS and BHT groups significantly improved both their BBS and BHT 1RM, suggesting a degree of transferability. However, the BBS group improved their BBS 1RM to a greater degree than the BHT group  and ILEX. Our data suggest that strength coaches, personal trainers, and trainees can self-select multi-joint lower-body trunk extension exercises based on preference or variety. However, evidence suggests that neither the BBS nor BHT exercises can meaningfully increase ILEX strength. Since strengthening these muscles might enhance physical and sporting performance we encourage strength coaches and personal trainers to prescribe ILEX exercise. Low-back strength, particularly as a component part of core strength and stability, retains importance in athletic performance and thus strength and conditioning within sports . Indeed,Research has supported that common exercises such as the barbell back squat (BBS) place considerable stress on the lumbar musculature  supporter = 0.93; Nevertheless, whilst the BBS exercise appears to be a hypothetical solution to strengthening the lumbar muscles , alternative and perhaps superior exercises may be available. An exercise growing in popularity is the barbell hip thrust (BHT), which has been shown to produce greater EMG amplitude in the hamstring and gluteal muscles when compared to the BBS exercise . The BHTHowever, the BBS and BHT both involve the lumbo-pelvic complex in a compound movement integrating both hip and lumbar extension (i.e. trunk extension). Previous research considering both resistance machines  and freeDespite this, a recent study  has repoTo date, no empirical studies have assessed the efficacy of either the BBS exercise or the BHT upon ILEX strength. Thus, the aim of the present study was to consider the efficacy of a 4-week mesocycle of either BBS or BHT exercise in increasing ILEX strength.n = 7) or BHT (n = 7) training. Both groups were assessed pre- and post-intervention for BBS and BHT maximal strength  as well as isometric ILEX strength. The study was approved by Solent University Health, Exercise, and Sport Science (HESS) ethics committee (ID No. 669).A randomised trial, research design was used whereby 14 trained males were randomised in to either BBS (n = 7), or BHT (n = 7). Subjects were asked to refrain from any exercise away from the supervised sessions. Participant demographics are included in Using convenience and snowball sampling methods, 14 trained males were recruited. Subjects were required to have >6 months resistance training (RT) experience, and currently be performing a structured RT programme with at least one session per week including the use of BBS exercise and BHT exercise and have no history of low-back pain. Written informed consent was obtained from all subjects prior to participation. Subjects were randomised using a computer randomisation programme to one of two groups; BBS (n). The ES was calculated for both studies using p < 0.05. It should be noted that the study was powered for the identification of within-group changes (two tailed) in outcomes and between group comparisons were a secondary outcome. Based upon sensitivity analysis for between group comparisons for the analysis detailed below the study was powered to identify at most a large between group ES of f = 0.82.A power analysis of previous research, using similar study design and asymptomatic subjects  was condr = 0.94–0.98; Subjects attended a preliminary session where they were assessed on their familiarity with both BBS and BHT exercises and verified their ability to perform them safely with correct technique. They also attended a familiarisation session for the MedX lumbar extension machine  used for measuring isometric ILEX, where subjects performed a testing session in the format detailed below. The ILEX machine  has beenSubjects reported to the laboratory having refrained from exercise other than that of daily living for at least 48 h before baseline testing, and at least 48 h before post-intervention testing. Maximum strength testing was consistent with recognised guidelines established by the NSCA . Prior tThe BBS 1RM was completed first, followed by the BHT 1RM, with a 20-min rest interval between exercises to allow for sufficient recovery. As per On a separate day following no less than 72 h rest, subjects attended the laboratory for ILEX strength testing. Subjects were seated upright with their pelvis secured by a restraint pad across the anterior, upper thigh, and another across the thigh just superior to the knee. These pads were fixed tightly to ensure the effort produced was from the lumbar musculature only and not from the pelvis or thighs, isolating the lumbar extensors. A counter-weight was used to balance the mass of the upper body and the effects of gravity on the upper body. All subjects were assessed for range of motion (ROM) and performed a dynamic warm-up with a load equating to 90 lbs/∼41 kg and three submaximal isometric tests at full flexion, full extension and a mid-range position. Maximal isometric testing was then performed at seven joint angles  where subjects were encouraged to gradually achieve maximal effort over 2–3 s and to maintain the maximal contraction for a further 1 s. The torque produced was measured by a load cell attached to the movement arm. Subjects rested for 5–10 s between tests at different joint angles. Isolated lumbar extension strength was considered as a ‘strength index’ (SI) calculated as the area under the torque curve (using the trapezoid formula) from multiple angle testing in order to provide a composite measure of overall changes in strength across the ROM see . Based uBoth the BBS and BHT participant groups attended two RT sessions per week, for a 4-week mesocycle. During these sessions each participant performed three sets using 80% of their 1RM  in a controlled, non-explosive manner  in order to maximise muscle tension and eliminate momentum. All sets were performed to momentary failure . Thus, for these variables, due to the significant deviations from normality of distribution combined with the relatively small sample size, the data were rank transformed prior to performing ANCOVA (The independent variable was the group (BBS or BHT) and dependent variables changes  in BBS 1RM, BHT 1RM, and SI. Analysis of covariance (ANCOVA) was used for between group comparisons in dependent variables with baseline measures as covariates . Point estimates were calculated along with the precision of those estimates using 95% CI for within-group adjusted means. The 95% CI were further interpreted to indicate that significant within-group changes occurred if the upper or lower limits do not cross zero. The software used was SPSS  and the cut off for significance was atistics  for datag ANCOVA . All resF = 5.240, p = 0.043), change in BHT 1RM  = 6.673, p = 0.025), but not for change in SI  = 1.541, p = 0.240). The 95% CI for changes did not cross zero for change in squat 1RM or change in BHT 1RM in either group and thus both groups had significant within-group improvements in these outcomes which also exceeded the typical errors of measure. The 95% CI for changes in SI did not cross zero for the BHT group suggesting a significant change; however, change in SI did not exceed typical error of measurement for either group and so is unlikely to be a meaningful change. Between group comparisons using ANCOVA revealed significant differences for change in squat 1RM , the present data suggest a degree of transferability between exercises. In considering transferability, it is noteworthy that the largest individual strength increase in BBS 1RM (25 kg) was a participant in the BHT group, and the largest individual increase in BHT 1RM (55 kg) was a participant in the BBS group. Whilst our data cannot explain individual differences based on the possible heterogeneity of the participant group, we might reflect on the degree of training status of these participants that accommodates such large strength increases over a 4-week training cycle. Any attempt at explaining these large individual changes would be purely speculative; however, we postulate that they might have arisen from the mesocycle prior to this intervention period not being focused on maximal strength, and/or that participants might have benefited from direct individual supervision, as has been seen in previous research with some training experience . As suchincreasing lumbar extension strength through isolated training might also increase performance. The absence of specific isolated lumbar extensor training may be suboptimal for developing athletic performance. Indeed, However, our data suggest that neither the BBS nor BHT RT exercises serve to improve ILEX strength, since the pre- to post-intervention changes did not exceed the typical error of measurement. Recent research serves to support our findings in context of the BBS exercise. For example, The BHT represents a more contemporary, and thus limited, area of exercise science research. Certainly evidence has suggested that the BHT might be an efficacious exercise for improving sprint performance and mid-thigh pull , and datWith the above in mind it appears that, despite large increases in BBS and BHT 1RM as a result of the respective RT programmes, ILEX strength is likely not improved by either exercise. Previous research has suggested that the use of exercise where pelvic rotation is permitted does not improve ILEX strength . In contr = 0.71–0.94) and vertical jump  and ILEX. Our data suggest that both the BBS and BHT exercises produce meaningful increases in strength which might transfer between lower-body trunk extension exercises. This allows strength coaches, personal trainers, and trainees to self-select multi-joint lower-body trunk extension exercises based on preference or variety. However, evidence suggests that neither the BBS nor BHT exercises, nor indeed any exercise allowing pelvic rotation through trunk extension, can meaningfully increase ILEX strength. Since strengthening these muscles might enhance physical and sporting performance we encourage strength coaches and personal trainers to supplement existing practices by prescribing specific ILEX exercise.10.7717/peerj.7337/supp-1Supplemental Information 1Click here for additional data file.10.7717/peerj.7337/supp-2Supplemental Information 2Click here for additional data file.


Findings from studies of prenatal exposure to pesticides and adverse birth outcomes have been equivocal so far. We examined prenatal exposure to agricultural pesticides in relation to preterm birth and term low birthweight, respectively, in children born between 1998 and 2010, randomly selected from California birth records. We estimated residential exposure to agriculturally applied pesticides within 2 km of residential addresses at birth by pregnancy trimester for 17 individual pesticides and three chemical classes . Among maternal addresses located within 2 km of any agricultural pesticide application, we identified 24,693 preterm and 220,297 term births, and 4412 term low birthweight and 194,732 term normal birthweight infants. First or second trimester exposure to individual pesticides  or exposure to 2 or more pesticides in the three chemical classes were associated with a small increase (3–7%) in risk for preterm birth; associations were stronger for female offspring. We did not find associations between term low birthweight and exposure to pesticides other than myclobutanil  and possibly the pyrethroids class. Our improved exposure assessment revealed that first and second trimester exposure to pesticides is associated with preterm delivery but is rarely linked with term low birthweight. During the first decade of the 21st century, the rates of preterm birth and low birthweight were 11–13% and 7–8% in the US, respectively ,5,6, andCalifornia is the largest agricultural state in the United States, with more than 150 million pesticide active ingredients applied every year . Pestici2 section [However, epidemiologic studies examining the effects of pesticides on preterm birth and low birthweight have yielded inconsistent results. While ecological and cross-sectional studies have reported positive associations between preterm birth and low birthweight and pesticide use in agriculture ,15,16, r section . The oth section  or pre-e section  for expoOur objective was to examine whether prenatal exposure to agricultural pesticides contributes to the risk of preterm birth or term low birth weight. We estimated GIS-derived exposure to agricultural pesticides applied near maternal residences during pregnancy, which were selected based on previous research that indicated reproductive toxicity, and we considered trimester-specific exposure windows in all agricultural regions of California .n = 339,210), as described previously [n = 143,595) [n = 20,124), with extreme or implausible gestational ages (<20 weeks or >45 weeks) or birthweights (<500 g or >6800 g) (n = 6390), with missing sex (n = 2), with home addresses outside of California (n = 1433), as well as multiple births  and also removed duplicate subjects. The remaining births included 41,089 preterm births, defined as having a gestational age less than 37 weeks, and 358,256 term births (not low birthweight) between 37 and 41 weeks 6 days as the reference. To examine term low birthweight, we used 7407 term births with a birthweight of less than 2500 g, indicating intrauterine growth restriction, and included 317,710 term normal birthweight infants (2500–4000 g) in the comparison group. We restricted our study population to those residing at birth within 2 km of fields on which agricultural use pesticides were applied, and we included 24,693 preterm births and 220,297 term births, and 4412 term low birthweight and 194,732 term normal birthweight infants. We combined two sets of birth records randomly selected from all California births between 1998 and 2010; these were (1) controls matched to children with autism in a 1:10 ratio by sex and birth year (eviously  and (2) 143,595) , which wWe geocoded maternal residential addresses listed on the birth certificates using an automated approach  and calchttp://www.pesticideinfo.org/) that have been linked to reproductive toxicity [We selected 17 individual chemicals previously observed to have reproductive toxicity ,37,38,39toxicity ,37,40,41Since the specific locations of non-agricultural pesticide applications  are not provided by the PUR, and due to competing exposures such as air pollution in urban areas ,43, we rWe conducted unconditional logistic regression analyses adjusting for matching factors (infant sex and year of birth) and the source of control subjects (autism vs. cancer study) to estimate the odds ratios (ORs) and 95% confidence intervals (CIs) of associations between pesticide exposure and preterm birth or term low birthweight. To account for the unbalanced sex ratio  and birth year distribution in this combined sample, we included the inverse of the sampling fraction  as a stabilized weighting factor to reflect the sex and birth year distribution of all California births. We additionally adjusted for potential confounders based on the literature ,45,46,47Infants born preterm or born term with a low birthweight were more likely to have mothers of younger age, lower education level, lower neighborhood SES, who started prenatal care after the first trimester, and who used Medi-Cal or another government program instead of private insurance. In addition, preterm births were more likely to be a third or later born child, and to have a mother with Hispanic or Black race/ethnic origin. Term low birthweight infants were more likely to be female and a first-born child, born to Black or Asian mothers .Exposure to pesticides in the first and second trimesters was associated with a small increase in risk for preterm birth (ORs: 1.03~1.07)  while th1st trimester: 1.04; 95% CI: 1.00–1.08), or pyrethroids  increased the OR for preterm birth in the group exposed to 2 or more chemicals in each class compared with no exposure  in infants exposed to two or more pyrethroids  of mothers move during pregnancy in the US . While mIn summary, this study found that first and second but not third trimester exposure to almost all pre-selected pesticides known or suspected to be reproductive toxicants was associated with preterm delivery, but only one pesticide (myclobutanil) and perhaps the pyrethroid class was related to term low birthweight. These associations were stronger in female infants, suggesting possible sex specificity for some of these agents or increased vulnerability in male fetuses that results in selective pregnancy loss.


Weight at birth is an important predictor of neonatal mortality and morbidity in dogs. In addition, the birthweight of the puppies in a litter influences the decision to perform a cesarean section. The goal of the present study was to estimate heritabilities for the puppy birth weight in Labrador retrievers.Of the 1138 Labrador retriever litters whelped at the Guiding Eye for the Blind between September 2001 and February 2018, 1013 were included in the analyses after data editing. Puppy weight at birth was the target trait, measured on a continuous scale in pounds, and converted to grams. Linear mixed models were used to identify factors influencing puppy weight at birth. The analyses showed that the sex of the puppy, litter size, length of gestation, adult weight of the dam, parity, year of birth and inbreeding coefficient of the puppies and dams contributed to the variance of the puppy birth weight. Dam and litter effects were included as random effects. A multiple trait derivative free restricted maximum likelihood approach was used to estimate variance components and genetic parameters with two animal models, one without covariates (Model 1) and one with covariates (Model 2). Sex of the puppy and litter size had moderate effects, whereas gestation length, adult weight of the dam, parity, year of birth and inbreeding coefficients of the dam and the puppies had minor effects. Estimates for Model 1 and Model 2 were 0.21 and 0.17 for the direct heritabilities, 0.22 and 0.22 for the maternal additive genetic heritabilities, 0.07 and 0.07 for the maternal permanent environmental proportions, and 0.14 and 0.08 for the environmental proportion of the litter.In order to estimate reliable breeding values for puppy weight at birth, sex of puppy, litter size, length of gestation and the adult weight of the dam should be included. Estimates could benefit from weighing the dams prior to each mating. The size or weight of puppies at birth is an important parameter, which influences neonatal mortality and morbidity and is a predictor of puppy survival –3. AfterFactors with an influence on the birthweight have been reviewed for domestic animals . Among tA low birthweight may be caused by a short gestation time and/or intrauterine growth retardation as reported in humans and other mammalian species including the dog , 9. PossBirth weight is not only determined by the genetic makeup of the offspring and its environment, but also by the maternal genetic composition and environment provided by the dam . EstimatGuiding Eyes for the Blind is a non-for profit organization that breeds and trains mostly LR and a few GS to provide guide dogs to people who are blind or have visual impairment. The breeding strategy and detailed procedures of Guiding Eyes for the Blind have been described . The aimp ≤ 0.05) for PBW and were included in model 2  than females (472 g). Adult dam weight, inbreeding coefficient of the dam and puppy, sex, year of birth of the litter, length of gestation, litter size, parity and parity squared were identified to be significant covariates  who wants to investigate birth weight in a specific population, canine or not, or improve the situation with respect to birth weight in specific populations. Results suggest that the inclusion of covariates in the model improves the estimates of variance components. The magnitude of the heritabilities indicates that estimation of breeding values could improve breeding program with respect to PWB. Whether our findings help to improve the situation with respect to the stabilization of PWB in guide dogs depends heavily on how they can be implemented in a breeding strategy that is focused on the guiding abilities of the dogs. To evaluate the impact of our results on the general LR population, reliable data on PWB need to be collected. Most of the covariates included in model 2 could also be recorded in the field.Statistical analyses were carried out using Stata/SE 15.1  and MTDFREML . The pedUsing MTDFREML two models were used to estimate variance components of PWB. Model 1 contained no covariates, because covariates are not always easily recorded in the field, and Model 2 included the covariates identified to influence PWB (Additional file Additional file 1. Effects of covariates.Additional file 2. Sex-specific bivariate analyses of male and female PWB.Additional file 3. Estimation of the dominance and sex-linked genetic variance components.Additional file 4. Datafile.Additional file 5. Descriptive statistics of variables.Additional file 6. Statistical Analyses of the birth weight in Labrador Retrievers using Stata.Additional file 7. Animal model of the birth weight in Labrador Retrievers using MTDFREML.


Stochastic birth–death models provide the foundation for studying and simulating evolutionary trees in phylodynamics. A curious feature of such models is that they exhibit fundamental symmetries when the birth and death rates are interchanged. In this article, we first provide intuitive reasons for these known transformational symmetries. We then show that these transformational symmetries  are preserved even when individuals at the present are sampled with some probability. However, these extended symmetries require the death rate parameter to sometimes take a negative value. In the last part of this article, we describe the relevance of these transformations and their application to computational phylodynamics, particularly to maximum likelihood and Bayesian inference methods, as well as to model selection. Linear birth–death models play a pivotal role in phylodynamics. These stochastic models provide a prior distribution on evolutionary trees (both the shape and edge length distribution) for Bayesian inference methods . MoreoveThe study of such models dates back to some classical papers from the early to mid-20th century ,b, and tIn this article, we identify and explore curious symmetries in fundamental birth–death model probability distributions when the birth and death rates \docThis equation states the surprising result that the probability of one individual having one surviving descendant after time Theorem 1.For any non-negative value of More generally, set http://dx.doi.org/10.5061/dryad.57704ft). To provide some intuitive insight into this result, we now provide a direct and conceptually transparent Proof of Theorem 1 in the case where Scenario 1: The process starts at time 0 and is stopped at time Scenario 2: The process starts at time This result has been established in Intuitively, the result of the time-reversed process with birth and death being interchanged is analogous to the forward-in-time process. However, we justify this intuition by a formal argument showing that the probability of observing one individual after time Consider some population size trajectory The probability density of For Now we reverse time in the realization Note that Analogously, One can directly extend this argument to establish Theorem 1 for any value of We continue to study a birth–death model with constant and non-negative birth and death rates Let us first suppose that we start with one individual at time 0, and let Theorem 2.For withFor the critical case withSupplementary Material available on Dryad.For ample in . For theIn what follows, we investigate the expressions for We introduce two new variables Note that when The following lemma is straightforward to verify using simple algebra .Lemma 3.For all can be written as functions of only two parameters  depend on all three parameters ment} by  and is tAs highlighted in Remark 6, we can, based on Corollary 1 of the supplementary material available on Dryad, directly conclude thatThus, we obtain the same probability density when swapping birth and death. As a consequence, we have to specify if the birth rate is bigger or smaller than the death rate prior to any analysis based on these equations.Supplementary Material available on Dryad.When using the tree probability densities in a maximum likelihood inference framework, the expressions are maximized over the parameters for a given tree. Based on the five conditioned tree probability density equations, we should optimize over ested in , SectionTheorem 7.Let Given Given In summary, given we estimate a negative Given the dependency of Supplementary Material available on Dryad.Next, we characterize all birth–death parameters that are transformations of Theorem 8.Let if Note that the parameters Corollary 9.With Next, we consider Corollary 10.With Properties of the birth–death tree distribution need to be known in order to test if empirical data are significantly different from these properties and thus the birth–death model has to be rejected for the given data. Sometimes, proofs of the properties of the conditioned tree distribution are carried out for complete sampling (i.e., for parameters For a given phylogenetic tree, it is tempting to ask if a model with Birth–death models have been studied for almost 100 years . HoweverSuch parameter nonidentifiability has important consequences for using birth–death models in phylogenetic and phylodynamic inference. In particular, the likelihood surface of the three birth–death parameters Furthermore, we showed that for some of the parameter triplets (The birth–death model presented here is the simplest model for speciation and extinction, or for transmission and recovery. However, it has limitations for explaining the data, as it assumes exponential growth of the population, although populations cannot have unlimited growth, and it assumes that all individuals are dynamically equivalent. There has been considerable work on extending the birth–death model to address such limitations , but no


Purpose: Health disparities persist in birth outcomes by mother's income, education, and race in the United States. Disadvantaged mothers may experience benefit from supplements to family income, such as the earned income tax credit (EITC). We examined the effects of state-level EITCs on birth outcomes among women with a high school education or less, stratified by race and ethnicity.Methods: A quasi-experimental multistate and multiyear difference-in-differences design is used to assess effects of the presence and generosity of 23 state-level EITC laws on birth outcomes from 1994 to 2013. The methods utilized the U.S. National Vital Statistics System birth data for the outcomes: birth weight, probability of low birth weight , and gestation weeks.Results: Across all subgroups, any level of state EITC is associated with better birth outcomes with the largest effects seen among states with more generous EITCs. Black mothers experience larger percentage point reductions in the probability of LBW and increases in gestation duration. Among mothers with a high school education or less, results translate into 3760 fewer LBW babies with black mothers and 8364 fewer LBW babies with white mothers per year at the most generous state EITC level . Hispanic and non-Hispanic mothers display relatively similar effects.Conclusions: The EITC at the federal and state level is an effective policy tool to reduce poverty and improve birth outcomes across racial and ethnic subgroups. Given the historically higher risk among black mothers, state-level EITC expansions offer one policy option to address this persistent health disparity. Across racial/ethnic groups, a clear graded association exists between income quintile and low birth weight (LBW).3 This association persists with maternal education. Mothers with a high school education or less are at significantly higher risk for having a LBW infant compared to mothers with some college or more.3 In 2016, LBW levels ranged from 7% for births to non-Hispanic white women to nearly 14% among those to non-Hispanic black women.4 Rates among Hispanic subgroups, which are more similar to rates among whites than blacks, ranged from 7% for births to Mexican women to 9.5% for births to Puerto Rican women. Evidence suggests that inequities in birth outcomes between black and white mothers are caused by a lifetime of differential exposure to risks, including income inequality, educational achievement gaps, residential segregation, and toxic environmental exposures.6 The current study examines whether a key policy strategy to improve family income among low- to middle-income working families—the earned income tax credit (EITC)—improves birth outcomes similarly or differentially across race and ethnic subgroups.There are striking and persistent health inequities in birth outcomes by mother's income, education level, and race in the United States. Disparities by race are caused by a complex set of social factors across the life course, two of the most important being opportunities for quality education and employment. Compared with the majority white population, blacks are more than twice as likely to be poor (22% vs. 9%) and are less likely to have a college degree (23% vs. 36%).7 There is consistent evidence that federal and state EITCs positively affect families' economic circumstances; increase participation in the labor force, particularly among single mothers; reduce poverty, including child poverty; improve educational outcomes among children; and improve health outcomes among mothers and children.8–10 For example, multiple studies examining the effects of the federal12 and state13–15 expansions of the EITC on birth outcomes have found significant associations between the implementation of EITC laws with decreases in LBW and increases in average birth weight, especially among those of lowest socioeconomic status and the greatest increase in EITC income. We recently conducted a longitudinal study of changes in amount and refundability of EITCs across 23 states from 1994 to 2013 and found improvements in infant health outcomes (both birth weight and gestation weeks) in states with EITCs, with larger effects seen among states with more generous EITCs.13 We found little difference in maternal health behaviors  associated with state-level EITCs.The U.S. federal EITC has been credited as the largest and most effective antipoverty program for families in the United States.9 one study reported an iatrogenic effect of the federal EITC expansions during the 1990s on infants born in California to non-Hispanic black women who were eligible for Medicaid and had a high school education or less.16 Specifically, 2 months after EITC disbursement, there was an increase in the odds of very LBW (<1500 g) among infants born to black women.16 However, this association was not significant at 3 or 4 months after EITC disbursement. A similar pattern was observed among births to non-Hispanic white women in California, although the effect was smaller. It is difficult to understand why the results of this study are inconsistent with the majority of the extant literature, which finds beneficial effects of EITCs.Given disparate birth outcomes by socioeconomic status and race, it is surprising that there is so little research on the effects of the EITC by race and ethnicity. In contrast to research documenting positive outcomes following federal and state EITC expansions,12 EITC effects were smaller among Hispanic mothers than non-Hispanic mothers, perhaps because Hispanic children tend to have better baseline birth outcomes and some do not qualify for the EITC as undocumented immigrants. They suggest that the differences by race may be largely attributed to the fact that black mothers are more at risk, having almost double the risk of low-birth weight births and lower income. Results of their study imply that the EITC may be a policy strategy that will not only contribute to reducing negative birth outcomes but also birth inequities.A more comprehensive study of the effects of federal expansions of the EITC during the 1990s examined effects across all states on similar “high impact” mothers (single women with high school education or less). The study results indicated that the federal EITC expansions reduced the likelihood of having a LBW baby (<2500 g) for black mothers by 0.73 percentage points (relative to a mean of 14.4%); this is more than four times larger than the effect on white mothers (0.13 percentage point decline relative to a mean of 8.1%).13 The current study expands upon our previous study by examining differential effectiveness of the EITC across race and ethnic subgroups. Black mothers are at greatest risk of low income, low educational attainment, and poor birth outcomes, and, therefore, may experience greater benefit from supplements to family income. Given the increased risk among black mothers and their babies, and the larger federal EITC effect among blacks found by Hoynes et al.,12 we hypothesize that the beneficial effect of a state EITC will be larger among black babies.We previously evaluated effects of the presence and generosity of 23 state-level EITC laws on maternal health behaviors and birth outcomes from 1994 to 2013 among women with a high school education or less. We found few statistically significant differences in maternal health behaviors associated with state-level EITC. By contrast, results for infant health outcomes of birth weight and gestation weeks consistently showed improvements in states with any level of EITC and larger effects within states with more generous EITCs.To ensure reliable and valid indicators of state policy changes from 1994 to 2013, we conducted original legal research on EITC policy for each of the 50 states, plus Washington DC. In collaboration with a team of legal researchers, we developed a codebook and detailed coding protocol to capture important EITC policy dimensions, including eligibility criteria and amount and refundability of the credit. Quality control procedures included blinded independent coding of a random sample of items by two trained legal researchers. All coders were closely supervised by a senior attorney, who reviewed protocols with coders for any variable showing 5% or higher cross-coder disagreement rate. All divergences between two coders were resolved by the supervising attorney after meeting with the two coders and examining the original legal text.If the state has an EITC, the value is usually expressed in the law as a percent of the federal EITC, but some states specify a dollar amount of the credit which we converted to a percent of federal based on the relevant federal dollar amount. The values of the state credits often vary based on number of children living in the household. In addition, some states specify that the EITC is refundable, meaning that if tax liability falls to zero, the government will send a refund check for the credit amount. Nonrefundable credits provide no further income beyond a zero tax liability.13 We use the 10% cutoff because this is the median value of EITC among states over the sample period. In 1994, 5 states had an EITC in place, but by 2013 this number grew to 26 states, plus Washington DC. Maryland is excluded from analysis because of the unique structure of their EITC law, which does not match the measurement model used for all other states. Our legal research identified 80 changes in state EITC law from 1994 to 2013. After condensing into our 5 policy categories, the 80 legal changes represented 34 shifts in policy category across 23 states over the study period. Variable operationalization details have been previously reported.13 Washington state enacted a refundable credit of 5% of the federal EITC in tax year 2009 and was scheduled to rise to 10% in 2010. Due to state budget shortfalls, policy makers have not yet financed the credit. Therefore, we placed Washington state into the no EITC policy category.We use the information gathered to create a series of indicators combining the presence of and generosity of EITC payments as follows: (1) states with no EITC (reference category); (2) states with an EITC, nonrefundable payments, and payments less than 10% of the federal amount; (3) states with an EITC, payments that are refundable, and payments less than 10% of the federal amount; (4) states with an EITC, nonrefundable payments, and payments 10% or more of the federal amount; and (5) states with an EITC, refundable payments, and payments 10% or more of the federal amount.13 We also limit the sample to singleton births, and we do not include young teenage mothers (women less than age 18).Infant health outcomes and maternal characteristics come from the U.S. National Vital Statistics System, which records a 100% census of U.S. births annually. Infant health outcomes include birth weight, probability of LBW (less than 2500 g), and gestation weeks. We cannot identify the individual women who assuredly qualify for the EITC, instead we use mother's education to limit the sample and estimate models that represent an approximation of intent-to-treat.Maternal characteristics recorded on birth certificates include: mother's age, marital status, mother's education, and mother's race/ethnicity. In addition, we use county geographic identifiers to merge county-level covariates potentially related to the outcomes. These include the county unemployment rate, real income per capita, percent poverty, number of obstetricians/gynecologists and primary care physicians per 1000 women ages 15–44, and county population size. Specific county characteristics are unavailable for counties of less than 100,000 people. For these small counties, we use the average value for all counties in the state with a population less than 100,000. All models also include state fixed effects and year-by-quarter fixed effects. Summary statistics by race  and ethnicity  for all variables included in the models are shown in Birth certificate data and state EITC data were merged based on the number of prior children and outcome specific theory driven lags. For maternal outcomes, we merged EITCs from the year before conception. For infant outcomes, we merged the EITC received closest to the time of birth.We estimated race and ethnicity specific effects of state EITCs using fully-interacted, fixed-effect panel data models. That is, all covariates described above enter the model after being multiplied by the individual-specific indicator variables for race (black and white) and, subsequently, by ethnicity (Hispanic and non-Hispanic). All models were estimated as linear probability models for ease of interpretation of the estimates. Models were re-estimated as logistic regressions without any substantive changes. All models accounted for within-state serial correlation by clustering the standard errors at the state level.The equality of the race and ethnic specific estimates for each of the EITC indicators was tested with a F-test of the interacted coefficients. We respecified the model to generate coefficients that reflect the additional impact of EITC for blacks (Hispanics) differing from the white (non-Hispanic) effect and conducted an omnibus F-test of this set of black (Hispanic) EITC coefficients. The results of these tests are reported in the text below and in the notes to the table.These fixed effects isolate the effect of state EITCs so that our EITC indicator estimates reflect the average difference before and after the change in each policy within states, after subtracting the before and after differences in states without a state-specific EITC, effectively treating states without a state-specific EITC as a control group. To bias our estimates of the effect of state-specific EITCs, potential confounders would need to change differentially over time between states, and these changes would need to occur at a similar time to changes in our state-specific EITC indicators. To protect against such confounding, we included the previously discussed individual-level and county-level covariates in all models. Birth certificate data does not allow us to ascertain which individual women qualify for the EITC. To better approximate and intent-to-treat analysis, we restrict all models to women who have reported a high school education or less.13 any level of state EITC is associated with improved birth outcomes, and the largest effects are seen among states with more generous EITCs (us EITCs .Birth weight gains are higher for black mothers in states with an EITC (16.12–37.16 g representing 0.5–1.2% increases in birth weight) than for white mothers (9.38–28.40 g representing 0.3–0.9% increases in birth weight), but were not significantly different between black compared with white mothers, except among states with a high EITC with no refund.p=0.0048) and gestation weeks (omnibus test p=0.0001). Reductions in the probability of LBW for black mothers in states with an EITC range from 0.6 to 1.4 percentage points (depending on size and refundability of the credit), which represent 5–11.7% reductions from the mean of 12%. Reductions in probability of LBW for white mothers in states with an EITC range from 0.2 to 0.7 percentage points, which represent 3.3–11.7% reductions in the mean of 6%. Among black mothers with a high school education or less, these results translate into 1611–3760 fewer babies born LBW to the 268,580, on average, singleton and nonteenage births per year in the United States. Among white mothers with a high school education or less, these results translate into 2390–8364 fewer babies born LBW to the 1,194,920, on average, singleton and nonteenage births per year.Results show a larger beneficial effect among black mothers compared with white mothers for the probability of LBW . Among black mothers in states with high-value EITCs, on average, gestation weeks increase by small fractions of a week, representing 0.38–0.46% longer gestation weeks. The comparable numbers among white mothers in states with high-value EITCs are 0.17–0.41% longer gestation weeks.Finally, average gestation week increases are also similar for Hispanic and non-Hispanic mothers, with a high-value refundable state EITC increasing average gestation weeks by small fractions of a week, representing 0.2–0.3% increases. In states with a high-value nonrefundable EITC, Hispanic mothers have a lower gestation week increase  compared with non-Hispanic mothers .12 including our own study of state EITCs,13 we find beneficial effects of the EITC on birth outcomes across race and ethnic subgroups. We find larger percentage-point effects for black mothers with a high school education or less; however, given their higher baseline rate, the relative percent reduction is similar among black and white mothers. In states with the most generous state EITCs, refundable and 10% or more of the federal, we find nearly 12% reductions in LBW births for black and white mothers. Among mothers with a high school education or less, this reduction translates to 3760 fewer babies born LBW with black mothers and 8364 fewer babies with white mothers per year across the United States. In addition, gestation time is slightly longer and birth weights are higher among babies with black mothers. We find few differences in effects by ethnicity, including an inconsistent pattern of effects on gestation time.Consistent with previous research,12 The federal EITC expansions resulted in smaller treatment effects among Hispanic than non-Hispanic mothers. Hoynes et al. attribute this finding to a lower baseline rate of low birth rate among Hispanic mothers, speculating that there was less room for improvement.12 In our study, we find few statistically significant differences by ethnicity, although patterns are consistent with the findings of the federal EITC.The direction and magnitude of these results by race are similar to those found for the federal expansions of the EITC, with larger absolute beneficial effects for black mothers given their higher baseline rate of risk.17 In summary, our results are consistent with previous research finding overall beneficial effects of the EITC on birth outcomes across race and ethnicity.We use a strong quasi-experimental design to study the effects of size and type of state-level EITCs on birth outcomes by race and ethnicity. This study design controls for time trends and state differences in birth outcomes, as well as other important covariates. A limitation is the inability to specifically identify individual women who assuredly qualify for the EITC—instead we use mother's education to limit the sample and estimate models that represent an approximation of intent-to-treat. Another limitation is that the observed results on LBW could be an underestimation of the differential effect, given that black women have a rate of stillbirth twice higher than white women, and stillbirths are excluded from our data on live births.The federal and state EITC is an effective policy tool to reduce poverty and improve birth outcomes across racial and ethnic subgroups. Given ongoing disparities, it remains essential to advance policy options that may help reduce these disparities and explore other policy options that may prevent and further mitigate the effects of social disadvantage across the life course for those mothers most at risk.