This cohort study analyzes the association between participation in the Women, Infants, and Children (WIC) program and birth outcomes among low-income pregnant women in the United States by gestational age at birth and maternal race/ethnicity. Is receipt of Special Supplemental Nutrition Program for Women, Infants, and Children benefits during pregnancy associated with preterm birth and infant mortality among low-income expectant mothers in the United States?In this cohort study of 11 148 261 pregnant women with Medicaid coverage between 2011 and 2017, the proportion who also received Women, Infants, and Children benefits during pregnancy decreased from 79.3% in 2011 to 67.9% in 2017. However, receipt of those benefits was associated with decreased preterm birth and infant mortality among low-income women.Findings from this study suggest that greater emphasis on supplemental nutrition benefits could reduce the burden of preterm birth and infant mortality among low-income expectant mothers, who represented 42.9% of all expectant mothers in the United States. Nearly 4 in 10 expectant mothers in the United States received Special Supplemental Nutrition Program for Women, Infants, and Children (WIC) benefits during pregnancy between 2011 and 2017. Despite public support for the program, empirical evidence of the success of the program varies substantially.To assess the association of WIC program participation during pregnancy by low-income expectant mothers covered by Medicaid with infant mortality by gestational age at birth and by maternal race/ethnicity in comparison with their counterparts who did not receive WIC benefits.This cohort study obtained data from January 1, 2011, to December 31, 2017, from US live birth certificates. Data were from 11 148 261 expectant mothers who delivered live births in states that have implemented the 2003 revision of the US live birth certificate and whose insurance coverage and receipt of WIC benefits were recorded on the birth certificates. Data analysis was performed from June 2019 to October 2019.Receipt of WIC benefits during pregnancy.The first outcome was gestational age at birth: extremely preterm (<28 weeks), very preterm (28-32 weeks), moderate-to-late preterm (32-37 weeks), and normal term (≥37 weeks) births. The second outcome was death within the first year of life.P < .001). The odds of preterm birth compared with normal term birth were lower among expectant mothers covered by Medicaid who received WIC benefits during pregnancy compared with their counterparts who did not receive WIC benefits during pregnancy . The odds of mortality within 1 year of birth were lower for infants whose mothers were covered by Medicaid and received WIC benefits during pregnancy compared with those who did not receive WIC benefits during pregnancy .Among the 11 148 261 expectant mothers who delivered live births between 2011 and 2017 and were covered by Medicaid during pregnancy, the modal age at delivery was 20 to 24 years, the predominant race/ethnicity was non-Hispanic white (4 257 790 [38.2%]), and 8 145 770 (73.1%) received WIC benefits during pregnancy. The proportion of expectant mothers covered by Medicaid who also received WIC benefits decreased from 2011 to 2017 (79.3% to 67.9%; This study found that receipt of WIC benefits among expectant mothers with Medicaid coverage was associated with lower risk of preterm birth and infant mortality. Second, women with longer pregnancies may experience better birth outcomes because of their greater gestational period rather than because of receipt of WIC benefits . Accounting for these possible biases, some studies found a much lower or no association between WIC and gestational age, birth weight, and infant mortality. In addition, other studies found WIC participation to be associated with improved birth outcomes for low-income minority women, but either less or no association was found for low-income white women.17Other studies question these conclusions of success and commonly note 2 possible limitations.18This national-based cohort study addresses this controversy by assessing the association of WIC participation with infant mortality  using data from 11 million birth certificates with 1-year mortality follow-up. We also estimate the association of WIC participation with infant mortality by maternal race/ethnicity. We believe the study provides timely and contemporary empirical evidence to policy makers about the WIC program, which, like other safety net programs, could face possibly substantial budget cuts in the foreseeable future.20 The Dartmouth College Committee for the Protection of Human Subjects determined that institutional review board review and informed consent were not required for the present study because the regulatory definition of human participant research did not apply. The data used were deidentified and publicly available. This study followed the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) reporting guideline.This cohort study analyzed US birth certificate data from January 1, 2011, to December 31, 2017, collected by the National Center for Health Statistics as part of the National Vital Statistics System.21 By 2016, all 50 states and the District of Columbia had implemented this revision.22 This study used data from 25 263 716 expectant mothers who had live births in states that followed the 2003 rule and for whom the type of insurance coverage and receipt of WIC benefits were recorded on the birth certificate , very preterm (28-32 weeks), moderate-to-late preterm (32-37 weeks), and normal (≥37 weeks) births. The second outcome was death within the first year of life among infants who were born alive.The primary variable of interest was receipt of WIC benefits during pregnancy. Sociodemographic characteristics of expectant mothers included their age at delivery , race/ethnicity , educational attainment , marital status (married or unmarried), and source of payment for the delivery . We considered source of payment for the delivery as the type of health insurance coverage for the expectant mother. Pregnancy history included lifetime number of pregnancies (gravida), births of viable offspring (para), and abortions (aborta). Receipt of prenatal care was categorized as either none or at least some prenatal care during pregnancy. Obstetric complications included prepregnancy diabetes, gestational diabetes, prepregnancy hypertension, gestational hypertension, and hypertension eclampsia. Method of delivery and final route included spontaneous, forceps, vacuum, and cesarean. Smoking frequency was categorized as 0 or 1 or more cigarettes per day 3 months before pregnancy and 0 or 1 or more cigarettes per day during pregnancy. The proportion of births with missing data on any of these covariates ranged between 0.0% and 3.5%  whose births were covered by Medicaid, (2) who received WIC benefits during pregnancy, and (3) whose births were covered by Medicaid and who also received WIC benefits during pregnancy. We tested for time trends by fitting a least-squares regression line of each proportion against year.2 test with Yates continuity correction. Fourth, we calculated the infant mortality rate among expectant mothers covered by Medicaid who either did or did not receive WIC benefits during pregnancy, and we similarly assessed the difference between these proportions. The infant mortality rate equaled the number of deaths within the first year of life among infants who were born alive per 1000 live births.23Third, we calculated the proportion of extremely, very, and moderate-to-late preterm births among expectant mothers covered by Medicaid who either did or did not receive WIC benefits during pregnancy. We assessed the differences between each pair of proportions using Pearson χFifth, we fit a multivariable ordinal logistic regression model to identify the association between receipt of WIC benefits during pregnancy and preterm birth among expectant mothers covered by Medicaid. The outcome of the model was gestational age at birth . Other covariates included year of birth, sociodemographic characteristics , pregnancy history (GPA), receipt of prenatal care, clinical risk factors, and cigarette smoking before and during pregnancy.Sixth, we fit a multivariable logistic regression model to identify the association between receipt of WIC benefits during pregnancy and infant mortality among expectant mothers covered by Medicaid. The outcome of the model was infant mortality, and the covariates were identical to the first model as well as the method of delivery and gestational age category.24 We considered a common subpopulation of expectant mothers with the following characteristics: primigravid and nulliparous , aged 25 to 29 years, non-Hispanic white, married, and nonsmoker before and during pregnancy. We varied receipt of WIC benefits during pregnancy.Seventh, to illustrate the population-level implication of receiving WIC benefits during pregnancy, we estimated the probabilities of extremely, very, and moderate-to-late premature births and infant mortality according to the results of the regression models in steps 5 and 6.Eighth, we fit similar regression models as those in steps 5 and 6 and included the interaction of race/ethnicity and receipt of WIC benefits during pregnancy to assess whether the associations differed by race/ethnicity. We also fit a multivariable logistic regression model to evaluate the association between receipt of WIC benefits during pregnancy and spontaneous birth compared with indicated birth by cesarean delivery among expectant mothers covered by Medicaid who delivered preterm births.P < .05. All statistical calculations were done with R, version 3.6.1 . We conducted data analysis from June 2019 to October 2019.Listwise deletion was used to remove cases with missing data in all regression models. Throughout the analysis, statistical significance was assessed at 2-sided The number of expectant mothers covered by Medicaid during pregnancy between 2011 and 2017 totaled 11 148 261 . The proportion of expectant mothers covered by Medicaid who also received WIC benefits during pregnancy decreased from 2011 to 2017 . The prevalence of moderate-to-late premature birth was 853 076 (10.5%) among those who received WIC benefits during pregnancy and 316 155 (11.2%) among those who did not (P < .001). The infant mortality rate was 5.2 deaths per 1000 live births among those who received WIC benefits during pregnancy and 8.2 deaths per 1000 live births among those who did not (a 36.6% relative risk reduction).Among expectant mothers covered by Medicaid, the prevalence of preterm birth was lower among those who received WIC benefits during pregnancy compared with those who did not. For example, the prevalence of extremely premature birth was 56 558 (0.7%) among those who received WIC benefits during pregnancy and 34 800 (1.2%) among those who did not  among mothers who did not receive WIC benefits during pregnancy, and the mortality rate was 5.98 deaths per 1000 live births  among mothers who received WIC benefits during pregnancy. This difference corresponded to an absolute risk reduction of 1.12 deaths per 1000 live births  and a relative risk reduction of 18.66% .The odds of preterm birth compared with normal gestational age birth were lower for non-Hispanic white , non-Hispanic black , and Hispanic  expectant mothers covered by Medicaid who received WIC benefits during pregnancy compared with their counterparts who did not receive WIC benefits during pregnancy  who also received WIC benefits decreased substantially over time. Second, receipt of WIC benefits during pregnancy was associated with reduced odds of preterm birth and infant mortality. Third, the magnitude of these associations was approximately equal among low-income non-Hispanic black, Hispanic, and non-Hispanic white women.28 This study adds to the existing literature by providing a contemporary estimate of the associations between WIC and preterm birth and between WIC and infant mortality.These findings are consistent with those of a large body of research that found participation in the WIC program was associated with improved birth outcomes.30 Second, WIC participation during pregnancy is associated with increased vitamin D intake, which may lower the risk of pregnancy-induced hypertension and preeclampsia .31 Similarly, WIC participation during pregnancy is associated with greater maternal iron intake, which may increase birth weight for gestational age.32 Third, the WIC program encourages breastfeeding by providing guidance, counseling, and breast pumps.33 Breastfeeding is associated with reduced risk of postneonatal death (death between 28 days and 1 year after birth).34Participation in WIC may lower the likelihood of preterm birth and reduce gestational age–specific infant mortality through several possible biological mechanisms. First, the food supplementation enabled by WIC participation is associated with higher overall and protein-specific caloric intake, both of which are associated with improved fetal growth and increased birth weight.13 assessed the association between WIC participation and 6 birth outcomes  and found no significant implications of the WIC program. However, assessment of WIC participation was based on maternal recall; for some participants, pregnancy occurred years or decades before they were surveyed. The present study was based on maternal recall of WIC participation at the time of delivery and was likely subject to less recall bias. Joyce et al12 noted that gestational age bias could have led to overestimated associations between WIC and birth outcomes. We accounted for gestational age bias and found reduced odds of infant mortality within each gestational age category.This study reaches substantively different conclusions from those of several studies that found no or little association between WIC benefits and key birth outcomes. For example, Foster et al15 Bitler and Currie5 argued that low-income women who enrolled in the program were more disadvantaged than their counterparts who did not enroll; however, the program was still associated with improved birth outcomes.3 Thus, any selection bias would lead to a conservative estimate of the association.5 We found that this pattern of relative disadvantage may persist; in the present study, low-income women enrolled in the WIC program during pregnancy had lower educational attainment compared with their counterparts who did not enroll in the program.Another potential limitation of studies on WIC and birth outcomes is selection bias. Some low-income women may have been more able to enroll in the WIC program and maintain and renew their benefits compared with other low-income women.17 For example, between 2005 and 2008, Khanani et al17 assessed the association between WIC participation and infant mortality in a single Ohio county  and found that the program was not associated with lower infant mortality among low-income white women. However, results from Hamilton County may not be generalizable to the rest of the United States. For example, WIC participation rates among pregnant women were lower , and infant mortality rates were higher .35 The present nationwide study found that receipt of WIC benefits during pregnancy was associated with lower infant mortality accounting for gestational age at birth, and this association was approximately equal in magnitude across race/ethnicity.Previous research found that the association between WIC and birth outcomes may not be equal across race/ethnicity.36 This decline may be associated in part with logistical barriers faced by low-income expectant mothers to enrolling and fully participating in the program.37 First, some women have reported long wait times for receiving benefits at WIC offices as a key barrier to program participation.38 Second, once women receive their benefits in the form of checks or vouchers, they may experience embarrassment or negative interactions with cashiers or other customers at stores when redeeming their benefits.39 Third, participants may face limited selection in shopping because some retailers may not carry WIC-eligible foods and products in allowable sizes and quantities.40 It is not known, however, whether these structural barriers have worsened over time. Nonetheless, the cumulative implications of these barriers may lead some low-income women to forgo WIC benefits in favor of Supplemental Nutrition Assistance Program benefits. The Supplemental Nutrition Assistance Program benefits are more generous  and disbursed through an electronic benefits transfer card rather than as a check or voucher (as with WIC).41The proportion of WIC-eligible pregnant women who actually received benefits decreased from approximately 59% in 2009 to 50% in 2016.42 To date, 35 states and the District of Columbia have expanded Medicaid by raising the income eligibility threshold.43 The income eligibility threshold varies by state: 17 states set eligibility up to 199% of the federal poverty level (FPL), 22 states set eligibility between 200% and 249% of the FPL, and 12 states and the District of Columbia set eligibility at ≥250% of the FPL; in 2019, the FPL was $21 330 for a family of 3.44 Actual WIC participation did not necessarily increase in states that raised the income eligibility for Medicaid after enactment of the Patient Protection and Affordable Care Act.45 Overall, national WIC participation among eligible pregnant women decreased from 56.8% in 2010 to 50.3% in 2016 despite Medicaid expansion during this period.46 California, for example, expanded Medicaid eligibility in 2010 and set income eligibility for pregnant women at 213% of the FPL. However, the number of pregnant women who received WIC benefits decreased from 187 000 in 2010 to 152 000 in 2016.1 Public health campaigns to increase enrollment and greater federal funding could ensure that all expectant mothers with low income or at risk for poor nutrition receive WIC benefits throughout their pregnancy.Pregnant women eligible for Medicaid are automatically eligible to receive WIC benefits. The Patient Protection and Affordable Care Act of 2010 enabled states to raise the income eligibility limit by providing federal government cost sharing to offset the cost of additional Medicaid beneficiaries.This study has several strengths. First, it used more than 11 million birth certificate records with linked mortality follow-up for 1 year. Second, it addressed a possible limitation of gestational age bias association with WIC and birth outcomes by directly assessing patterns in infant mortality and accounting for gestational age at birth.47This study also has several limitations. First, states adopted the 2003 revision of the US Standard Certificate of Live Birth, which ascertains health insurance coverage and receipt of WIC benefits for women during pregnancy, in different years. The temporal analysis based on this revision included 83% of all live births in 2011, 90% of all live births in 2014, and 100% of all live births in 2016 and 2017. Second, birth certificate data did not identify the frequency of WIC benefits use during pregnancy. Some expectant mothers may not have been able to redeem WIC benefits every month because of logistical barriers. Thus, we may conservatively estimate the association between WIC benefits and gestational age and infant mortality. Third, this study focused on live births and excluded pregnancies that resulted in miscarriage and stillbirth. Participation in the WIC program may reduce the incidence of these adverse outcomes among high-risk women in part through increased nutrition and prenatal visits.This cohort study found that receipt of WIC benefits during pregnancy was associated with lower preterm birth and infant mortality for low-income women. Promoting WIC enrollment through public health campaigns and increasing federal funding for the program could raise the number of expectant mothers with low income or at risk for poor nutrition receiving the benefits throughout their pregnancy.


To analyse subdural haemorrhage (SDH) during infancy in Sweden by incidence, SDH category, diagnostic distribution, age, co-morbidity, mortality, and maternal and perinatal risk factors; and its association with accidents and diagnosis of abuse.th version (ICD10), retrieved from the National Patient Register and linked to the Medical Birth Register and the Death Cause Register. Outcome measures were: 1) Incidence and distribution, 2) co-morbidity, 3) fall accidents by SDH category, 4) risk factors for all SDHs in the two age groups, 0–6 and 7–365 days, and for ICD10 SDH subgroups: S06.5 (traumatic SDH), I62.0 (acute nontraumatic), SDH and abuse diagnosis.A Swedish population-based register study comprising infants born between 1997 and 2014, 0–1 years of age, diagnosed with SDH-diagnoses according to the . Median age was 2·5 months. For infants older than one week, the median age was 3·5 months. Case fatality was 6·5%. Male sex was overrepresented for all SDH subgroups. Accidental falls were reported in 1/3 of the cases. One-fourth occurred within 0–6 days, having a perinatal risk profile. For infants aged 7–365 days, acute nontraumatic SDH was associated with multiple birth, preterm birth, and small-for-gestational age. Fourteen percent also had an abuse diagnosis, having increased odds of being born preterm, and being small-for-gestational age.Incidence of SDH was 16·5 per 100 000 infants (The incidence was in the range previously reported. SDH among newborns was associated with difficult birth and neonatal morbidity. Acute nontraumatic SDH and SDH with abuse diagnosis had similar perinatal risk profiles. The increased odds for acute nontraumatic SDH in twins, preterm births, neonatal convulsions or small-for-gestational age indicate a perinatal vulnerability for SDH beyond 1st week of life. The association between prematurity/small-for-gestational age and abuse diagnosis is intriguing and not easily understood. Subdural haemorrhage (SDH) is reported to have an incidence of 21·0–24·1 per 100 000 infants  and meniSDH occurs more often in infants than in older children . In a stCurrent knowledge about infancy SDH is based upon hospital studies, however, only two of these used codes from the International Classification of Diseases (ICD) to retrieve cases , 8. We hn = 395 812), thus having one or more diagnoses registered according to the Swedish version of the International Classification of Diagnosis (ICD-10) . Fr182 974) . For eacRegister . The finFor this study we included all cases of SDH according to their ICD-10 classification: SDH due to birth injury (P10.0); SDH-Acute nontraumatic subdural haemorrhage (I62.0); and SDH-Traumatic subdural haemorrhage (S06.5) . The difWe determined:Incidence of all included SDH and of the two subgroups S06.5 and I62.0, in addition to SDH combined with abuse diagnosis.Risk factors for SDH 0–6 days, all SDH 7–365 days, then separately for the subgroups S06.5, I62.0, for those infants who had both diagnoses (S06.5 and I62.0), and, finally, for SDHs combined with abuse diagnosis at 0–365 days.Co-morbidity for all SDHs except birth-related, only S06.5 and only I62.0.Population incidences were calculated as cases per 100 000. Mantel-Haenszel Chi-Square or Fisher´s exact tests were used to assess differences in non-parametric variables, means and analysis of variance (ANOVA). Crude and adjusted odds ratio (OR) analyses with 95% confidence intervals were applied to assess possible associations between maternal and perinatal exposures and the outcome. The statistical software package IBM SPSS 25·0  was used for data analyses.The Regional Ethical Review Board in Uppsala approved the study (2014-11-19 No 383).In all, 306 cases of infant SDH were found during the years 1997–2014, which gives an overall incidence of 16·5 per 100 000 infants. The distribution by diagnosis is shown in When split between the diagnostic subgroups, the incidence for each of these was 7·5 for I62.0, 5·2 for S06.5, 3·0 for P10.0, and 0·8 for the combination of I62.0 and S06.5.A total of 43 (14%) had a combination of abuse and SDH diagnoses; of these, 19 had S06.5, 15 had I62.0, and nine had both. The incidence of SDH with abuse diagnosis was 2·3 per 100 000 infants.Case-fatality rate was 6·2% (19/306). The causes of death were: malformations (5 infants), perinatal causes (1 infant), Menkes disease (1 infant), leukaemia (1 infant), subarachnoid haemorrhage (1 infant), unspecified acute lower respiratory infection (1 infant), peritonitis (1 infant), accidents (3 infants), specified accident events undetermined intent (3 infants), abuse/homicide (2 infants). Concomitant diagnoses were: intracerebral haemorrhage, traumatic brain oedema, cerebral infarction, skull fracture, sepsis, heart disease, cardiac arrest, and coagulopathy .Of the five infants who died from accidents or abuse/homicide, two died at the age of 38 and 355 days without any diagnosis registered prior to death. For the other three, the following diagnoses had been registered prior to death: anoxic brain injury due to drowning after a boat accident, unspecified anaemia with failure to thrive, and extremely low birth weight in combination with acute pharyngo-laryngitis and severe anaemia.p < .001). This gender preponderance was also present and stable when the material was split into diagnostic SDH subgroups   and 108 were girls (35·3%) (d I62.0)  and mont7th, 8th ; or betwOne-fourth of the cases, 74 (24%), occurred during the first week of life, 40 (13%) occurred after the age of 6 months. Mean and median age at diagnosis for all SDH cases was 3·3 and 2·5 months, respectively . For theAn accidental fall at time of diagnosis was reported for 104 infants (34·0%), 67 boys (64·4%) and 37 girls (35·6%), with a mean age of 122.7 days; nine occurred during first week of life. The distribution of the fall cases by age and type of fall are shown in A transport accident was reported at time of diagnosis for 39 infants (12·7%), with a mean age of 3·3 months. Even among the transport accidents, there was a marked male preponderance: 25 (64·1%) boys, and 14 (35·9%) girls. One infant with SDH and abuse diagnosis had a transport accident 1 day before diagnosis.p .001); the second-most common diagnosis was convulsions– 31 (12·4%)–without any differences between the diagnostic subgroups. Retinal haemorrhage was diagnosed in 27 (10·8%) infants, of these infants, I62.0 was more common than S06.5 (< p .05), Eighteen of these infants also had an abuse diagnosis  had distinct characteristics associated with dystocic labour [OR 2·74 (95% CI 1·68–4·46)], emergency caesarean delivery, assisted vaginal delivery [OR 18·8 (95%CI 4·60–78·8)], asphyxia, preterm birth, birth injury to the scalp or skeleton, and neonatal morbidity, especially convulsions [OR 271 (95%CI 171–431)] and other cerebral disturbances of the newborn [OR 117 (95% CI 50·9–272)] . Mode ofAdjusted odds ratios (aOR) are shown in This national, ICD-10-based study comprising 306 infants shows an incidence of infant SDH of 16·5 per 100 000 infants. One-fourth occurred during the first week of life; these cases had distinct birth risk profiles of difficult birth, preterm birth, and neonatal morbidity. Male sex, multiple birth, preterm birth, small-for-gestational age, and neonatal convulsions had increased odds for SDH at 7–365 days, especially for acute nontraumatic SDH—I62.0. Light or moderate traumas  were reported in 1/3 of the cases. One in seven– 14%–of all SDH cases had a concomitant diagnosis of abuse; these had, surprisingly, an even higher perinatal risk profile than the I62.0 subgroup, with a strong association between prematurity/small-for-gestational age and abuse diagnosis; moreover, those with abuse diagnosis also had a higher number of diagnoses of retinal haemorrhage, rib fractures, and long bone fractures.Throughout the entire set of material, there was a marked male preponderance, indicating that infant boys somehow are twice as prone to encounter a SDH as girls are, regardless of the diagnostic or age subgrouping of the cases. Boys have previously been shown to be overrepresented in very early publications on infant SDHs , 5. SimiIn this context, it is of interest that, in 1953, Guthkelch had already discussed the possibility that the male SDH preponderance could be explained by a birth trauma, stating: “the preponderance of males might be explained on the basis that male infants have bigger heads than females, and therefore more likely to sustain intracranial injury at birth.” , 21, 26.In 1944, Ingraham and Matson had already observed in their pioneering article that SDH in infants mainly occurred during the first months of infancy , 27–29.Subdural blood collections, most asymptomatic, occur in vaginal birth in very high percentages –33; almoNumerous reports have demonstrated that BEH in infants may predispose for spontaneous, nontraumatic subdural bleedings , 34–38, Even for the SDH with a traumatic aetiology, and for transport accidents analysed separately, there was a marked male preponderance. For infants, one would expect such incidents to be equally distributed among the genders; when it is not, one may wonder whether there is some underlying biological factor that renders boys more vulnerable to even trivial external trauma, as discussed above.In our study, very few co-morbidity diagnoses were registered, whereas clinical studies have reported much higher proportions of convulsions, vomiting, and irritability , 4, 6; oPrematurity and small-for-gestational age are distinct findings of increased risk of SDH-Acute non-traumatic and SDH with abuse diagnosis. Few studies have addressed these associations. Prematurity has been reported for SDH with abuse diagnosis by Keenan et al. , but notOne strength of the present study is that the data were retrieved from national registers, thus covering the entire country and, consequently, we included all infants who were given an SDH diagnosis throughout the study period. Another strength is that the same ICD-10 classification was used during the entire study period. The dataset including the referents contained 49% of all children born in Sweden during the study period, strengthening its population representativeness. Regarding exposure, a biological association is strengthened by the gradient of odds for SDH shown for preterm birth and small-for-gestational age. The validity of the Swedish health registers is considered to be high, both in regards to the Medical Birth Register  and the This lack of validation for the SDH diagnoses is a limitation, as we had no access to detailed clinical documentation on each patient’s history, condition, neuroimaging, ophthalmological examinations, or treatment. Neither could we assess the initiation time point of the SDH. In addition, we have no validation of the registered diagnoses of co-morbidity, as we have had to rely completely on the hospital physicians’ choice of ICD-10 classifications. This is a major clinical dilemma, as there are no available SDH diagnoses in ICD-10 for common clinical entities, such as “chronic subdural haematoma”, “subacute blood collections”, “subdural hygroma”, or “subdural effusion” , or any Another weakness relates to the registering of traumatic SDHs. We know that parents suspected of violent shaking or intended head injury can be asked to give an explanation for an SDH that is assumed to be of traumatic origin; in that situation, they may give emphasis to minor traumas during the preceding days, such as bumping the head into a bed-post or a short fall onto a sofa, as reported in the UK and Ireland in 41/97 of the SDH cases diagnosed as abuse [The SDH incidence found in this study was in the range previously reported, although lower than that reported in a study from the British Isles over a one-year data collection period in 1998–99 [The case fatality rate in our study was lower; 6·2%, compared to the 19% reported by Hobbs et al. in 1998–1999 [Our study identified 18% birth-related SDH, a finding similar to the 14% reported by Hobbs et al. , 44.Fall accidents were reported for 104 infants, all classified as traumatic. In the British Isles, only 7 out of 189 cases were classified as accidents, although carers provided a history of minor trauma in 41 of the 97 non-accidental head injury (NAHI) cases , which aIn our study, one in seven cases had a diagnosis of abuse, 2·3 per 100 000, while more than half in the UK and Republic of Ireland were classified as NAHI, at 14·2 per 100 000 . This diThe present study reflects the shortcomings of the ICD-10 system in classifying SDH; there is no code for chronic or subacute haematomas, nor is there a code for subdural fluid collections containing small amounts of blood or no blood at all, or what in clinical terms are described as “hygromas”. In the present ICD version, any of these chronic conditions therefore must be categorized into one of the two available SDH diagnoses (I62.0 or S06.5), both indicating that the SDH was caused by an acute internal or external event. In the diagnostic work-up of suspected violent shaking, this lack of diagnostic differentiation may represent a serious pitfall, as even obviously chronic subdural collections have to be classified as an acutely acquired condition.The vulnerability of boys in attaining SDH, whether traumatic or not in the first months of life, is indeed intriguing and may reflect the possibility that at least some of these male infants are predisposed to subdural blood collections with no or only minor trauma. External hydrocephalus is such a condition; it occurs mainly in boys and during the first months of life, and it predisposes for subdural fluid collections. The fact that perinatal risk factors, such as being prematurely born or being small-for-gestational age, are highly associated with an SDH-related abuse diagnosis underscores the suggestion that the diagnostic work-up should be cautious and should include consideration about whether external hydrocephalus might be an underlying cause of the SDH.As no routine head scan is performed on all newborn infants, there is a high risk that subtle SDH related to perinatal risk factors will not be detected postnatally. For this reason, a future investigation of the long-term development of infants with postnatal subdural blood, using non-radiation head scans, is warranted.This study cannot fully assess whether there has been an over-diagnosis of shaken baby syndrome or abusive head trauma; however, the similar risk profiles for infants with SDH and infants with SDH and abuse diagnosis indicate that the diagnostic process might be uncertain. Thus, the insufficient scientific evidence of the triad in identifying traumatic shaking, as shown in a systematic literature review of the Swedish Agency for Health Technology Assessment and Assessment of Social Service , raises This population study should be followed by a clinical study with access to information from clinical records, patient history, and imaging, such as MRI scans.The total incidence of SDH was in the range previously reported– 16·5 per 100 000 infants—with the majority of cases occurring during the first months of life (median 2·6). Boys were twice as prone to having an SDH-diagnosis across all ages and SDH subgroups, indicating an increased vulnerability to minor trauma and/or increased bleeding sensitivity in male infants. SDH among neonates was associated with birth and neonatal morbidity. Traumatic SDH during infancy was strongly related to fall accidents. Acute nontraumatic SDH and SDH with abuse diagnosis during infancy had similar perinatal risk profiles. To be able to correctly identify all causes of intracranial bleeding, population-based newborn cohorts should undergo repeated imaging studies from birth to several months of age.S1 Tableth revision of the International Statistical Classification of Diseases (ICD-10).Swedish version of 10(DOCX)Click here for additional data file.S2 TableN = 908,565). ANOVA (means) comparing all and by gender: 1) S06.5 with I62.0, 2) I62.0 with SDH and abuse diagnosis. Mantel-Haenszel Chi-Square or Fisher exact. P-value  comparing: 1) all SDH with population, 2) S06.5 with I62.0, 3) I62.0 with SDH and abuse diagnosis.Source population: children born in Sweden: the National Patient Register and the Swedish Medical Birth Register, Swedish National Board of Health and Welfare ((DOCX)Click here for additional data file.


Results showed that erdite could synthesised in nanorod form with a diameter of approximately 200 nm and lengths of 0.5–3 μm by adding MnO2; moreover, the crystals grew with increasing MnO2 addition. Without MnO2, erdite particles were generated in irregular form. The capacity of the erdite nanorods for tetracycline (TC) adsorption was 2613.3 mg/g, which is higher than those of irregular erdite and other reported adsorbents. The major adsorption mechanism of the crystals involves a coordinating reaction between the −NH2 group of TC and the hydroxyl group of Fe oxyhydroxide produced from erdite hydrolysis. To the best of our knowledge, this study is the first to synthesise erdite nanorods and use them in TC adsorption. Erdite nanorods may be developed as a new material in the treatment of TC-containing wastewater.Erdite is a rare sulphide mineral found in mafic and alkaline rocks. Only weakly crystallised fibrous erdite has been artificially synthesised via evaporation or the hydrothermal method, and the process generally requires 1–3 days and large amounts of energy to complete. In this study, well-crystallised erdite nanorods were produced within 3 h by using MnO Erdite is stable in alkaline conditions and can be observed in the residual sludge obtained from alkaline desulfurisation using sodium ferrite as desulfuriser during the recycling of lead from electrolyte battery waste2 and from aluminate salt purification during bauxite refining3. Erdite was first synthesised by melting a stoichiometric ratio of Na, S and Fe salts, but the product was intertwined in the form of weakly crystallised fibres4. Similar erdite fibres were produced by heating siderite and NaHS at 150 °C, but the hydrothermal process required 2–3 days to complete5. Crystallised erdite can be used as raw material in the water, chemical and semiconductor industries7; therefore, developing new, cost-effective and efficient methods to mass produce the material is highly desirable.Erdite is naturally associated with sulphide minerals in mafic and alkali host rocks2S and sulphur via the protonation of S-bridges and subsequent redox reaction of Fe(III) –SH groups8. Lassin et al.2 investigated the dissolution–precipitation of lead-refining slag (containing 20–50% erdite) from waste battery disposal and found that the final solution is alkaline; in addition, the predominant species were Na+, Fe(OH)4−, HS− and SO42−. The breakdown of erdite may generate fresh Fe oxyhydroxide, which has a large number of surface coordination sites and could be used to adsorb pollutants in wastewater9. However, the removal of pollutants in wastewater by using erdite has yet to be studied.Under acidic and/or neutral pH conditions, erdite is metastable and rapidly decomposes to produce H10. Because TC is poorly metabolised, large fractions of the drug are excreted through urine and faeces as the unmodified parent compound11. In conventional biological wastewater treatment, TC is mainly removed by biosorption, but only small portions of it are biodegraded13. The discharge limit and acceptable level of TC have not been published14. To avoid TC pollution, Chinese government strengthened regulations to limit the chemical oxygen demand of the treated effluent to less than 50 mg/L15. TC forms residues and is frequently detected in soil and surface waters near the discharge outlet of pharmaceutical wastewater treatment plants17. Many approaches, including wet oxidation, electrochemical treatment, photocatalysis18, adsorption20, coagulation and chlorination, have been developed to effectively remove TC from wastewater. Advanced oxidation employs free radicals to nonselectively oxidise organic matter (e.g. TC)22 without generating secondary pollutants23; its efficiency can be improved by enhancing the photocatalyst activity and/or reaction conditions26, but its applications are limited by high energy consumption and complicated instrumentation27. As an alternative to this method, adsorption is characterised by high effectiveness, cost efficiency and simply operation28; thus, it can be applied to treat TC-bearing wastewater. Various natural minerals and synthesised materials, such as montmorillonite29, activated carbons30, active sludge12, graphene31 and Fe–Mn binary oxide32, have been used to remove TC from wastewater. The adsorption performance of sorbents can be considerably improved by increasing the availability of surface sites and/or activating surface functional groups. Thus, the development of efficient adsorbents for TC removal remains an urgent undertaking.Pharmaceutical tetracycline (TC) is used worldwide in food additives, human therapy and farming industry2 as an auxiliary reactant. Using the proposed method, erdite nanorods were generated within only 3 h. The synthesised erdite nanorods showed high capacity for TC adsorption and, therefore, great application potential in wastewater treatment.In this study, we synthesised well-crystallised erdite nanorods via a one-step hydrothermal method using MnO2, the prepared P0 presents as irregular flake-shaped flocs. After MnO2 addition, nanorod particles with a diameter of approximately 200 nm and lengths of 0.5–2 μm are formed  and X-ray powder diffraction (XRD) . Figure med Fig. . The wea2 promotes microcrystal growth along the (110) reflection to produce the long erdite nanorods of P0.2. A peak at 2θ = 23.1°, which belongs to S8 (JCPDS 78-1888), was observed in all three spectra, thereby indicating that the polysulphide S8 is generated in the formation of erdite.When the Mn/Fe molar ratio was increased from 0.05 to 0.2, crystallisation was improved, and the length of the nanorod crystals increased to 3 μm Fig. . Moreove8 . The S 2p spectra in Fig. lts Fig. , which m2S is added to the liquid phase, it hydrolyses to HS− and OH−, resulting in increases in pH. When the pH of the solution is greater than 13, Fe3+ is hydrolysed to Fe(OH)4− 35. HS− is diffused into the structure of Fe(OH)4−, and OH− is replaced by HS− with the formation of Fe(OH)3HS− nn− being the final product (Eq. (2)nn− chain are neutralised by Na+ 36, and free channels within these chains are occupied by water molecules37. Therefore, well-crystalised erdite consisting of long linear (FeS2)nn− chains, with each Fe ion tetrahedrally surrounded by four sulphur atoms38, is generated.When Na1 and R2 represent the growing polymeric species of the (FeS2)− unit.In Eq. −and polymerisation of (FeS2)nn− nn− chain is stable in structure, even after calcination at approximately 500 °C for 12 h39. This finding indicates that the redox reaction of Fe–S in erdite is considerably inhibited.During erdite formation, replacement of OHnn− 4− and polymerisation of (FeS2)nn− chains, is produced as a result of the three reactions 4−, which inhibits erdite formation.Therefore, a large amount of OHons , qe is the equilibrium adsorption capacity (mg/g), KL and qm are the Langmuir isotherm constant and maximum adsorption capacity (mg/g), respectively, KF and bF are the Freundlich constants, KDR is the D–R isotherm constant and A and B are the Temkin isotherm constants.The adsorption capacity of the erdite nanorods for TC was investigated because the latter is a common pollutant in pharmaceutical wastewateruir (Eq. ), Dubiniuir (Eq. ) and Temuir (Eq. ) isotherR2) and the results are shown in Table The linear fittings of the four isotherm models are shown in Fig. 2 is important to the adsorption capacity of erdite for TC. MnO2 is a common auxiliary reagent and widely used to promote the performance of adsorbents toward various contaminant (e.g. TC)47. For instance, Song et al. reported that graphene loaded with 40% nanorod MnO2 exhibits a maximum TC adsorption capacity (qm) of 1789 mg/g and that the observed adsorption capacity remains constant with increasing MnO2 loadig from 40% to 60%31. In our study, P0.2 showed a desirable TC qm of 2613.3 mg/g, which is higher than those of P0, 40% MnO2-loaded graphene31, carbon materials (maximum of 1340.8 mg/g)30 and zeolite materials (800 mg/g)19. P0.2 also showed a TC removal rate higher than those of other flocculants nn− chains are disintegrated via the reverse reaction of Eq.  could be generated by addition of MnO2, which is subsequently hydrolysed to generate more Fe oxyhydroxide and MnOOH  was mixed with 0.01 g of the erdite particles in an Erlenmeyer flask. The flasks were sealed with parafilm and shaken at 200 rpm in an incubator  at room temperature. After 24 h, the flasks were taken from the incubator, and the erdite particles were separated by agitation at 5500 rpm for 5 min. The TC concentration in the supernatant was determined by high-performance liquid chromatography  using a mobile phase composed of 0.1% phosphoric acid and methanol at a ratio of 60:40 (v/v). The absorbance at 268 nm was read by using a UV detector, and the retention time was approximately 5 min. The adsorption capacity  of erdite was calculated by using the following equation:C0 and Ce are the initial and equilibrium concentrations of TC (mg/L), respectively, V is the solution volume (L), and m is the particle dosage (g).TC, a typical antibiotic found at high levels (above 920 mg/L) in pharmaceutical wastewater−9 mbar. Narrow-region electron spectra were acquired with an analyser pass energy of 20 eV. Binding energies were calibrated against that of the C1s peak (284.6 eV). The fitting curves were obtained using Thermo Avantage software  with a Shirley background and a Gaussian–Lorentzian peak model.The morphology of the erdite particles was determined by field-emission scanning electron microscopy . The X-ray diffraction (XRD) patterns of the erdite particles were analysed by a diffractometer  with Cu–Kα radiation in the 2θ range of 10°–40°. X-ray photoelectron spectroscopy (XPS) spectra were obtained using an X-ray photoelectron spectrometer  operated at 150 W with monochromatised Al–Kα X-rays (hν = 1486.6 eV). The base pressure in the analytical chamber was ≈102 as an auxiliary reactant. Addition of MnO2 considerably promoted the formation of erdite nanorods through the efficient generation of NaOH from the disintegration of Na2S. The generated erdite nanorods exhibited a TC adsorption capacity higher than those of previously reported adsorbents. The formation of stable –NH3+ groups between the hydroxyl group of Fe oxyhydroxide obtained from erdite hydrolysis and the –NH2 group in the TC side chain is the major TC adsorption mechanism of erdite.Erdite nanorods were synthesised via a facile hydrothermal method using MnO


In developed countries, prenatal maternal stress has been associated with poor fetal growth, however this has not been evaluated in rural sub-Saharan Africa. We evaluated the effect of prenatal maternal stress on fetal growth and birth outcomes in rural Ghana.Leveraging a prospective, rural Ghanaian birth cohort, we ascertained prenatal maternal negative life events, categorized scores as 0-2 , 3-5 (moderate), and > 5 (high) among 353 pregnant women in the Kintampo North Municipality and Kintampo South District located within the middle belt of Ghana. We employed linear regression to determine associations between prenatal maternal stress and infant birth weight, head circumference, and length. We additionally examined associations between prenatal maternal stress and adverse birth outcome, including low birth weight, small for gestational age, or stillbirth. Effect modification by infant sex was examined.β = − 0.91, p = 0.04; p-value for trend = 0.04). Among girls, moderate and high prenatal maternal stress was associated with reduced birth weight  and head circumference . In girls, high prenatal stress increased odds of any adverse birth outcome . Sex-specific analyses did not demonstrate significant effects in boys.In all children, high prenatal maternal stress was associated with reduced birth length .Date of Registration: April 14, 2011.Status of Registration: Completed. Low birth weight, a global marker of intrauterine growth, is an independent predictor of early-neonatal, late-neonatal and post-neonatal infant mortality, even after accounting for gestational age at delivery . Head ciIn higher-income countries, maternal psychosocial stress during or prior to pregnancy has been associated with low birth weight, prematurity, intrauterine growth retardation and other negative birth outcomes –10. To oSex-specific effects of maternal prenatal stress on birth outcomes have also been reported. Prior work suggests that female fetuses differentially adapt to intrauterine stressors by decreasing growth rates. A study in Israel of chronic maternal stress found that female fetuses were at increased risk for preterm birth and low birth weight . SimilarWe therefore leveraged a prospective pregnancy cohort to examine the impact of prenatal maternal stress as indexed by negative life events (NLEs) on infant birth anthropometrics and birth outcomes in rural Ghana. We hypothesized that infants born to women with higher levels of prenatal stress would have impaired birth anthropometrics and increased risk of adverse birth outcomes.The study was conducted in the Kintampo North Municipality and Kintampo South District located within the middle belt of Ghana. Subjects were recruited from the Ghana Randomized Air Pollution and Health Study (GRAPHS), which has been described elsewhere . BrieflyThe analysis in this manuscript include 353 pregnant women enrolled in GRAPHS who were in the third trimester of pregnancy between July 2014 to June 2015, and after informed consent underwent an additional maternal stress assessment in the third trimester of pregnancy. All enrolled participants had birth anthropometrics available for analysis. Procedures were approved by human studies and institutional ethics committees at Kintampo Health Research Centre (KHRC), the Icahn School of Medicine at Mount Sinai and the Columbia University Mailman School of Public Health and written consent was obtained from all study participants.Stress theory postulates that when environmental demands exceed coping strategies, we experience stress and consequent physiological disruption. Vulnerability to stress and physiological disruption is increased when events are experienced across multiple domains. Prenatal maternal stress was measured in the third trimester of pregnancy using the Crisis in Family Systems-Revised (CRISYS-R) survey, suitable for use in lower-income settings . The CRIAll birth anthropometrics were measured within 24 h of birth. Birth weight was measured to the nearest gram using the Tanita BD 585 digital baby scale  after standardizing with a 1-kg (kg) weight. Birth length and head circumference were also measured to the nearest 0.1 cm (cm) using a standard measuring board and a Lasso-o™ respectively. Low birth weight (LBW) was defined as less than 2500 g. Gestational age at delivery was determined using ultrasound-established dates . Small-fMaternal age at enrollment, marital status, ethnicity and family home ownership were obtained through questionnaires at GRAPHS enrollment. Maternal height to the nearest 0.1 cm and weight to the nearest 0.1 kg were recorded at enrollment. Infant sex was recorded at delivery.As above, the CRYSIS-R assesses maternal stress across 11 domains; a negative life events (NLE) domain score is created by summing the number of domains with at least one reported negative event. The NLE score was categorized into three categories  around the study-determined interquartile range, or 0-2, 3-5, and > 5 respectively. First, we used univariate and multivariable linear regression models to examine associations between prenatal maternal stress and birth anthropometrics. Models were adjusted for covariates linked to stress and birth anthropometrics in prior research, including infant sex, gestational age at delivery, ethnicity, maternal age, weight, height, marital status, and socioeconomic status as measured by family home ownership. We additionally adjusted for GRAPHS cluster, as post-randomization differences in socioeconomic status between groups were noted, and study cluster was directly related to cookstove use . Effect modification by infant sex was examined in stratified analyses and by fitting interaction terms. Trend tests were performed.We then employed univariate and multivariable logistic regression to determine associations between prenatal maternal stress and adverse birth outcomes, including LBW, SGA, and a composite variable of infants born with LBW or SGA or stillbirth. For these analyses, the moderate (NLE 3-5) and high (NLE > 5) prenatal maternal stress groups were collapsed together to maintain adequate cell sizes. Models were adjusted for infant sex, ethnicity, cluster, maternal age, weight, height and marital status and socioeconomic status. A sensitivity model additionally adjusted for gestational age at delivery. Effect modification by infant sex was again examined in stratified analyses and by fitting interaction terms.p-value was less than 0.05. In subgroup analyses, interaction terms were considered suggestive of an interaction if the p-interaction was less than 0.10. Analyses were performed in R version 3.3.3 .Main effects were considered statistically significant if the N = 286 (81%)], relationship [N = 236 (67%)], and home issue [N = 247 (70%)] domains  with a significant trend test (p-for-trend = 0.04).Table β = − 0.16, p = 0.02; NLE > 5 β = − 0.18, p = 0.04) while no association was seen in boys (p-interaction = 0.009). Similarly, among girls, moderate and high prenatal maternal stress were associated with reduced head circumference  while no association was seen in boys (p-interaction = 0.006). An association was seen between high maternal prenatal stress and birth length in girls  but not boys (p-interaction 0.84). Among girls, the trend test suggested an exposure-response relationship for all outcomes .Table Table Our prospective study adds to the literature supporting a relationship between prenatal maternal stress and both infant birth anthropometrics and adverse pregnancy outcomes, suggesting that these relationships operate not only in higher-income countries but also in rural lower- and middle-income countries (LMICs), including Ghana. Specifically, our data demonstrate an exposure-response relationship between increased prenatal maternal stress and reduced birth length in all infants and an exposure-response relationship between increased prenatal maternal stress and reduced birth weight and head circumference in girls. Further, girls exposed to high levels of prenatal maternal stress were at increased likelihood of having an adverse birth outcome.Impairments in birth anthropometrics predict infant mortality, which are particularly prevalent in LMICs. While genetics certainly play a role, the importance of non-chemical prenatal environmental exposures is increasingly apparent . Our stuβ-hydroxysteroid dehydrogenase (11 β-HSD) enzymes and placental glucocorticoid receptor (GR) expression, and may be overwhelmed resulting in increased fetal cortisol exposure. Sheep exposed to a late gestation increase in cortisol have a significant reduction in biometric measurements of size [In the study population as a whole, our data suggest that increased prenatal maternal stress is associated with reduced birth length. The placenta, the maternal-fetal interface, buffers the impact of maternal environmental exposures to promote fetal development. Maternal stress increases maternal cortisol levels and induces inflammation, impacting placental activity . The pla of size . Animal  of size , 32. Inc of size –36.β-HSD2 [β-HSD1 [β-HSD2 activity and were associated with increased cord blood cortisol [β-HSD2 activity and increased cord blood cortisol in female fetuses, resulting in reduced fetal growth [β-HSD2 and increases in 11 β-HSD1 activity have been demonstrated in intrauterine growth-restricted female placentas [Our data suggest that girls are particularly vulnerable to the effects of prenatal maternal stress. The placenta is sexually dimorphic and may influence the impact of maternal stress differently in male versus female fetuses . For exaβ-HSD2 , increas [β-HSD1 , and alt [β-HSD1  resultin [β-HSD1 . In femacortisol . A studyl growth . Furtherlacentas .The finding that prenatal maternal stress is associated with impaired birth anthropometrics may have health effects beyond the perinatal period and may influence infant and adult health. Support for this hypothesis, known as DOHaD, originated from observation that fetal growth restriction correlated with subsequent high adult mortality, including cardiovascular, pulmonary, oncologic and cerebrovascular mortality . HoweverStrengths of this study include the prospective study design, non-smoking pregnant women and prenatally assessed maternal stress across multiple domains. Further, despite the rural LMIC nature of the cohort, we confirmed gestational age at enrollment via ultrasound and measured birth anthropometrics within 24 h of birth regardless of location of delivery . These data are the first in rural sub-Saharan Africa to elucidate the relationship between prenatal maternal stress and birth outcomes and build upon previous reports from higher-income countries linking maternal stress and birth outcomes. If replicated, given the importance of these birth outcomes in predicting infant mortality, our findings suggest new targets for public health intervention. Other strengths include the focus on a rural, LMIC population more likely to be impacted by perinatal mortality and impaired birth anthropometrics and our well-characterized cohort that allowed us to adjust for a number of important confounders and pathway variables.We also acknowledge limitations. Chronic maternal stress over the prior 6 months was assessed across 11 domains, but it is possible that additional stressors were present and not evaluated. Temporal associations to identify sensitive windows of exposure are unable to be explored, as we did not serially measure maternal stress across gestation. While we adjust for many important confounders, we did not have data on paternal height or co-varying environmental exposures such as maternal diet. Our study does not include biomarkers of maternal stress or placental analyses therefore we can only hypothesize potential mechanistic pathways. Future studies would be strengthened by the incorporation of biomarkers of both maternal and fetal stress response systems, including HPA axis function. It also would be useful to have additional longitudinal follow up, to determine the childhood health ramifications of prenatal maternal stress in this population.In summary, these data demonstrate that all infants exposed to prenatal maternal stress are at increased risk for decreased birth length and that girls are particularly vulnerable to the effects of maternal stress on birth weight and head circumference, with increased risk of adverse birth outcomes. Given the direct links between birth anthropometrics and perinatal mortality, specifically in LMICs, and morbidity and mortality across the life course, these data begin to elucidate at-risk groups and highlight potential future public health interventions. Counseling and validation of the importance of managing maternal stress during pregnancy may be a simple intervention that can raise awareness to the perinatal health effects of maternal stress.Additional file 1: Table S1. Reported negative life events (NLEs) domains of CRYSIS-R survey. Table S2. Associations between prenatal negative life events (NLEs) and adverse birth outcomes stratified by sex: Sensitivity model additionally adjusting for gestational age


This study assesses the prevalence of Vietnamese children receiving the hepatitis B (HepB) vaccine birth dose and explores its associated socioeconomic factors. We used the data of the Multiple Indicator Cluster Survey, 2014. We estimated the overall percentage of HepB birth dose vaccination among 0–23-month-old children and its percentages according to selected characteristics. Multiple logistic regression was applied.st quintile of mother's socioeconomic status (38.6%). Receiving HepB vaccine birth dose was associated with child's birth weight, mother's age, mother's education, socioeconomic status, and ethnicity. 62.8% of children received the HepB vaccine birth dose. The prevalence rates by selected factors ranged from 35.3% to 76.7%. The categories with the lowest prevalence rates were children who had low birth weight (41.6%), had a mother aged less than 20 years (35.3%), had a mother with primary or less education (42.7%), belonged to ethnic minorities (30.3%), resided in rural areas (59.9%), and were in the 1 This study identified vulnerable groups, upon which policy-makers should focus their efforts to equitably and sustainably tackle birth dose HepB vaccine coverage as well as the full vaccination coverage, thereby promoting long-lasting herd immunity in this country. Hepatitis B (HepB) is an infection of the liver caused by the HepB virus that attacks liver cells . The HepThe Western Pacific region is well-known as an area with a very high burden of HepB, accounting for nearly half of all chronic HepB cases worldwide , 4. VietThe HepB birth dose is a monovalent vaccine containing surface proteins of the HepB virus absorbed by aluminum hydroxide or monophosphoryl lipid A adjuvant . SurfaceIn Vietnam, the Kinh is the main ethnic group, and other ethnic groups account for up to 14% of the national population . More thRecognizing the neglected aspects of the HepB birth dose coverage across the country may contribute to developing policy to expand the prevention interventions for mother-to-child transmission of HepB virus. Exploring key socioeconomic factors that are as the barriers to the HepB birth dose vaccination could pave pathway for scaling up other specific healthcare interventions and therefore could contribute to investigation, prevention, and management of healthcare-associated HepB infections among both mothers and newborns in the community. In the present paper, we used the most updated data from the Multiple Indicators Cluster Surveys (MICS) of the United Nations International Children's Emergency Fund (UNICEF) in the survey round of 2014 to assess the prevalence of Vietnamese children who received the birth dose of HepB vaccination and to identify socioeconomic factors associated with receipt of the birth dose.Data from the MICS 2014 was used for this study. The MICS was conducted by the General Statistics Office in collaboration with the Ministry of Health and the Ministry of Labor, Invalids and Social Affairs. Financial and technical supports for the survey were provided by the United Nations Children's Fund and the United Nations Population Fund. The MICS is nationally representative survey covering a broad range of issues affecting the health, development, and living conditions of Vietnamese women and children. The number of children aged 0-23 months included in the 2014 MICS was 1382 .The main outcome variable in our study was a binary variable specifying whether a 0–23-month-old child received the first dose of HepB vaccination within 24 hours after birth. Data for the available were extracted retrospectively from the MICS 2014. In these surveys, data regarding the HepB birth dose vaccination were obtained from vaccination cards. If no vaccination card was available, the interviewers would ask mothers whether their child received a vaccination against HepB first dose vaccination , 16.The explanatory variables are as follows: child's sex , low birth weight defined as less than 2,500 grams (yes/no), mother's age (<20/20-35/36-49), mother's education (primary or less/lower secondary/upper secondary and higher), ethnic group (Kinh/Hoa ethnicity and minority ethnic group), living area , and mother's socioeconomic status.Mother's socioeconomic status (known as household's socioeconomic status) was measured as an asset-based wealth index and was constructed using principal component analysis (PCA). The MICS dataset included the Household Wealth Index which was calculated by the GSO of Vietnam. The index was based on the ownership of consumer goods, dwelling characteristics, water and sanitation, and other characteristics related to household wealth. Weights (factor scores) were assigned to correspond with individual household assets . The detBoth descriptive and analytical methods were used in the present paper. We estimated the overall percentage of HepB birth dose vaccination among 0–23-month-old children and the percentage according to sex, region, area, ethnicity, mother's education, and socioeconomic status. The Wald's Chi-square test was applied to compare the differences of receiving the HepB dose vaccination within 24 hours after birth among groups. The multiple logistic regression was used for the binary primary outcome variable to explore the factors associated with HepB birth dose vaccination. The cumulative probability of being vaccinated among ethnicities at age t among ethnic groups was estimated by the inverse Kaplan–Meier survival function (or 1−SKM(t)), also known to measure the fraction of children receiving the birth dose HepB vaccination for a certain amount of time . All staThis study was conducted on secondary data from the MICS with all identifiable information removed. The survey had obtained informed consent from the mothers before administering survey questionnaires and the consenting process had stated that data can be analyzed in subsequent analyses without retaking informed consent. All information in the original dataset was collected confidentially.st quintile mother's wealth status (38.6%). The difference in prevalence rates among all categories of each factor were significant  for most selected socioeconomic factors (except for the children gender).Prevalence rates and 95% confidence intervals for receiving the birth dose of HepB vaccination in each category by the selected socioeconomic factors are presented in The two curves of cumulative proportion for children receiving the birth dose of HepB vaccination belonging to Kinh/Hoa ethnicity and belonging to minority by children age in days are shown in Compared with the prevalence of receiving the birth dose of HepB vaccination among children with low birth weight, the odds was twice higher for the children with normal birth weight . Kinh/Hoa children had significantly higher odds of HepB birth dose vaccination than individuals from ethnic minorities . Mother's age was significantly associated with increased prevalence of receiving the HepB birth dose vaccination for their children. Children whose mothers had higher education were significantly more likely to have had completion of the HepB birth dose vaccination compared with those had mothers experiencing primary or less education. The odds of HepB birth dose vaccination were higher in children who belonged to the families with better economic status .In a Kate Whitford et al.'s report 2018), most immunization schedule in all studies included a first dose of HepB vaccine within 24 hours after birth except for one from Italy , most im. In VietIn our study, low birth weight children had significantly lower vaccination rates for HepB within 24 hours after birth. Saari et al. (2003) reported that vaccination rates could be affected by the infant's birth weight . InfantsIn our study, the prevalence of receiving HepB vaccine dose within 24 hours after birth was significantly higher in urban areas than in rural areas. Previous studies in Vietnam and elsewhere also found that urban areas had higher coverage of full immunization , 29. ThiIn Vietnam, there is a fact that women under the age of 20, who are in attending-school age, are not able to get married and give birth later. On the other hand, a finding of Linh Cu Le et al. reported that a risk of unintended pregnancy was 1.5 times higher in Vietnamese women marrying before 20 than those later or early marriage associated with unintended pregnancy . TherefoThe findings from this study have meaningful policy implications. This is the first study that assessed the coverage of timely HepB vaccine birth dose according to socioeconomic factors using the MICS data, which are the national representative immunization survey data in Vietnam. Importantly, the resource mobilization for the immunization in Vietnam is currently limited, not meeting the demand of expanded immunization . In the However, we acknowledge some limitations to this study. First, due to the limitation of the cross-sectional study design of the MICS, the results should be interpreted with caution so that they are not interpreted as implying causality. Second, the estimations of receiving HepB vaccine birth dose were derived from available information on vaccination cards and/or reported by mothers. Therefore, some children may have been vaccinated but did not have immunization card and were excluded from the study. In addition, mothers may have forgotten to report the vaccination of their children during their interviews. These would underestimate the birth dose of HepB vaccination coverage in this study. Finally, the cultural aspects related to uptake of HepB birth dose vaccination could not be assessed, such as acceptability and attitudes of Vietnamese women towards the vaccine quality.We found that the prevalence of receiving the first dose of HepB vaccine within 24 hours after birth did not meet the target of Vietnam (the immunization coverage of HepB birth dose should reach at least 65%). It is important to continue to coordinate with the WHO to support the implementation of newborn HepB vaccination; nevertheless, the targeted interventions in vulnerable population groups including child's low birth weight, mother's age less than 20, mother's low education, mother's low socioeconomic status, and child's ethnicity should be prioritized. There was a significant gap in the HepB vaccine birth dose coverage between the Kinh/Hoa ethnicity and minority ethnic groups, suggesting a need to improve both access and demand for HepB vaccine after birth among the other ethnic groups, in particular for minority ethnic groups living in remote and poor conditions. Furthermore, the policy developments in the HepB birth dose vaccination in particular and recommended vaccinations of the EPI in general should be established based on vulnerable populations, which leads to the sustainable interventions to decrease risk for healthcare-associated infections.


To determine the effect of intramuscular progesterone on the vaginal immune response of pregnant women with a history of prior preterm birth.A prospective, cohort study of women at 11–16 weeks gestation, ≥18 years of age, and carrying a singleton pregnancy was conducted from June 2016 to August 2017 after IRB approval. Women in the progesterone arm had a history of preterm birth and received weekly intramuscular 17-hydroxyprogesterone caproate. Controls comprised of women with healthy, uncomplicated pregnancies. Excluded were women with vaginitis, diabetes mellitus, hypertension, or other chronic diseases affecting the immune response. A vaginal wash was performed at enrollment, at 26–28 weeks, and at 35–36 weeks gestation. Samples underwent semi-quantitative detection of human inflammatory markers. Immunofluorescence pixel density data was analyzed and a P value <0.05 was considered significant.There were 39 women included, 10 with a prior preterm birth and 29 controls. The baseline demographics and pregnancy outcomes for both groups were similar in age, parity, race, BMI, gestational age at delivery, mode of delivery, and birth weight. Enrollment cytokines in women with a prior preterm birth, including IL-1 alpha , IL-1 beta , IL-2 , and IL-13  were significantly elevated compared to controls. In the third trimester the cytokine densities for IL-1 alpha , IL-1 beta , IL-2 , and IL-13  were all statistically similar between the progesterone arm and controls, respectively.There is an increased cytokine presence in vaginal washings of women at risk for preterm birth which appears to be modified following the administration of 17- hydroxyprogesterone caproate to levels similar to healthy controls. Intramuscular progesterone, administered to women at risk of preterm birth, reduces the likelihood of a subsequent preterm birth by approximately one-third ,2. The eAt the onset of labor, the ratio of progesterone receptor-A (PR-A) to progesterone receptor-B (PR-B) increases. Due to an increase in myometrial PR-A, there is a functional withdrawal of progesterone and an increase in sensitivity to contractile stimuli . This efRecently, Monsanto et al compared women with cervical insufficiency with normal controls and demonstrated elevated levels of proinflammatory cytokines in cervicovaginal fluid, suggesting a dysregulation of the local vaginal immune environment . PlacemeThe objective of this study was to evaluate the vaginal immune response of progesterone supplementation in pregnant women at risk for preterm birth compared to normal healthy pregnant controls.A prospective cohort study was conducted at an academic medical center from June 2016 to August 2017 with Stony Brook University Institutional Review Board (IRB) approval (CORHIS #2016-3441-F). Women were included if they were at a gestational age of 11 to 16 weeks, 18 years of age or older, and carrying a singleton pregnancy without acute or chronic vaginitis, diabetes mellitus, hypertension, and other chronic diseases or therapies that would affect the inflammatory or immune response. Women in the progesterone arm had a history of preterm birth and therefore were candidates for intramuscular 17-hydroxyprogesterone caproate (17-OHP). Controls were comprised of women with healthy uncomplicated pregnancies who did not receive progesterone therapy. All women were recruited prior to initiation of progesterone treatment. Intramuscular progesterone treatment was started at 16–20 weeks gestation in women with a prior preterm birth.A vaginal rinse was performed at three points in the study: on enrollment into the study at 11 to 16 weeks, at 26 to 28 weeks, and at 35 to 36 weeks gestation. All vaginal rinse samples were collected in subjects while in the supine position with a sterile speculum. A 10 ml syringe filled with 7 ml of sterile water attached to a soft plastic 14-gauge catheter was used to irrigate the vaginal vault. A sterile cotton swab was used to gently rub the mucosal surfaces. The fluid was aspirated back into a sterile 15 ml test tube. The samples were placed on ice and processed within 2 hours [The specimen tube was centrifuged at 3,000 RPM for 10 minutes. Aliquots of the vaginal lavage supernatant were stored at –80 degrees centigrade. The aliquots were later analyzed using Ray-Bio Human Inflammation Antibody Array C3 8-well plates  according to manufacturer’s instructions. Samples underwent semi-quantitative detection of 40 human inflammatory markers, and all cytokines were evaluated in duplicates. After treatment with biotinylated antibodies, followed by HRP-Streptavidin-labeled antibodies, the chemiluminescence detection of the products was performed using X-ray films. Numerical densitometry data were extracted by NIH software program Image Jv1.51j8 , using a dot array analysis plugin, as described in a previous publication .Patient demographics including age, parity, ethnicity, pre-pregnancy body mass index (BMI), and tobacco use, were extracted from the medical record. Pregnancy outcomes collected included gestational age at delivery, mode of delivery, birth weight, Apgar scores and Neonatal Intensive Care Unit (NICU) admission.Categorical variables were analyzed using, Chi square, Fisher’s exact test and continuous, normally distributed variables were analyzed using student’s t test. Cytokine data were normalized by dividing samples’ pixel density by pixel density of a positive control ,11. A P Thirty-nine women were included in the evaluation: 10 women with a previous preterm birth utilizing progesterone and 29 healthy control women. The baseline demographics and pregnancy outcomes of both groups were similar in age, parity, race, body mass index (BMI), gestational age at delivery, mode of delivery, and birth weight.  Overall,On initial evaluation at 11 to 16 weeks (visit 1), there were significant differences between inflammatory markers pixel densities in women with a history of preterm birth (progesterone arm) compared with healthy controls. Cytokines from women with a preterm birth including interleukin-1 alpha , interleukin-1 beta , interleukin-2 , and interleukin-13  were significantly elevated compared to controls . Evaluation of women in the progesterone arm in the early third trimester, 26 to 28 weeks (visit 2), demonstrated persistence of significantly elevated interleukin-1 beta  and interleukin-13  when compared with healthy women. Interestingly, interleukin-1 alpha , and interleukin-2  were statistically similar in both groups. . In the later third trimester, 35 to 36 weeks (visit 3), the cytokine densities for interleukin-1 alpha , interleukin-1 beta , interleukin-2 , and interleukin-13  were all statistically similar between the progesterone arm and the healthy controls, respectively.  , macrophage inflammatory protein-1 alpha  , eotaxin-1 , and monocyte chemoattractant protein-1 (MCP-1)  showed a significant downregulation when compared to controls.  On enrolThere is an increased cytokine presence in vaginal washings in women at risk for preterm birth which changes following the use of 17-OHP. There is a paucity of human investigations surrounding 17-OHP use and the possible mechanisms of action in the reduction in preterm birth in at risk women. Our work focused on theimmune inflammatory changes in the vaginal environment in women utilizing 17-OHP. Results demonstrate an initial increased presence of interleukin-1 alpha, interleukin-1 beta, interleukin-2, and interleukin-13 in women with a prior preterm birth compared to controls. Interleukin-1 was the first pro-inflammatory cytokine to be associated with infection-mediated spontaneous preterm birth –16. ImseIn women at risk for preterm birth, the densities for eotaxin-1, MCP-1 and MIP-1 alpha cytokines were significantly suppressed when compared with healthy women. Eotaxin-1, which is also known as C-C motif chemokine-11, has been associated with recruitment of eosinophils by inducing their chemotaxis, and is involved in allergic responses . Kraus eOne limitation in our study was the small sample size for both arms of the investigation; however, given the significance of the condition, we believe that this cohort still provides valuable evidence. Our unique approach to investigating the immune response following progesterone use did not allow for a pre-investigation sample size calculation, which is similar to other comparable investigations . AlthougThe precise mechanism of preterm labor is unknown and it is likely multifactorial. The pathophysiology which allows progesterone to work is also poorly understood. Yet, the use of progesterone in the prevention of preterm birth has been effective in the subpopulation of women at risk for recurrence. Based on our work, progesterone likely prevents preterm birth through alterations in the maternal immune system. We demonstrated pro-inflammatory cytokines, interleukin-1 beta, interleukin-1 alpha, interleukin-2, and interleukin-13, densities were significantly elevated in women with a prior preterm birth compared with controls. This was followed by a downregulation of these cytokines following weekly administration of intramuscular progesterone. Additionally, other cytokines, eotaxin-1, MCP-1 and MIP-1 alpha, were also reduced following progesterone therapy which have been implicated in the preterm labor process. Further investigation into the immunological changes resulting from progesterone use in the prevention of preterm birth is warranted.S1 TableThe cytokine pixel density percentage and standard deviation (STD) comparing preterm birth group and controls. Control women pixel density percentages are compared between visit 1 & 2 and visit 1 & 3 (Control Visits). Preterm birth group women pixel density percentages are compared between visit 1 & 2 and visit 1 & 3 (Progesterone Visits). Red shading signifies P<0.05.(XLSX)Click here for additional data file.