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BACKGROUND Both short and long interpregnancy intervals (IPIs) have been associated with adverse birth outcomes. We undertook a multistate study to describe the prevalence of selected birth defects by IPI. METHODS We obtained data from nine population-based state birth defects registries for singleton live births in 2000-2009 among mothers with a previous live birth identified through birth certificates. IPI was calculated as the difference between prior birthdate and start of the current pregnancy (conception date). We estimated prevalence of selected defects per 10,000 live births and prevalence ratios (PRs) with 95% confidence intervals (CIs) overall and stratified by maternal age at previous birth and race/ethnicity. Primary analyses focused on short IPI

BACKGROUND: The accuracy of birth outcome data provided by Internet-based cohort study participants has not been well studied. METHODS: We compared self-reported data on birth characteristics in Pregnancy Study Online (PRESTO), an Internet-based prospective cohort study of North American pregnancy planners, with birth certificate data. At enrollment, participants were aged 21–45 years, attempting conception, and not using fertility treatment. Women completed online questionnaires during preconception, early and late pregnancy, and postpartum. We requested birth certificate data during 2014–2019 from seven health departments in states with the most participants. After restricting to singleton births, we assessed specificity, sensitivity, and agreement comparing self-reported data from postpartum questionnaires with birth certificate data for gestational age at delivery (GA) and birthweight (grams). Our primary measure of self-reported GA (weeks) was calculated as: [280-(due date-birth date)]/7. We used log-binomial regression to assess predictors of agreement. RESULTS: We linked 85% (771/909) of women in selected states. Median age of women was 30 years (range: 21–42), 84% had ≥16 years of education, nearly 96% were married, 12% had household incomes

BACKGROUND: Arkansas (AR) had the highest prevalence of gastroschisis in a recent study including 15 U.S. states. Our objective was to evaluate trends in prevalence and the spatiotemporal distribution of gastroschisis in AR. METHODS: Infants with gastroschisis, born 1998–2015, were identified from the Arkansas Reproductive Health Monitoring System. Birth record data were used as denominators for calculations. Maternal residence at delivery was geocoded for spatial analyses. Annual prevalence rates (PRs) were calculated. Joinpoint regression analysis was performed to examine trends in gastroschisis and report the annual percent changes (APCs) in PRs. Spatiotemporal analyses identified counties with unusually high PRs of gastroschisis. Poisson regression, including county, year, and county*year indicators, was fit to evaluate the PRs of gastroschisis, while adjusting for county-level maternal variables. RESULTS: We identified 401 cases of gastroschisis among 694,459 live births. The overall PR of gastroschisis was 5.8/10,000 live births. The prevalence of gastroschisis had a significant APC of +5.3% (p < .0001) between 1998 and 2012, followed by a nonsignificant yearly average decrease of −17% through 2015 (p = 0.2). The Emerging Hot Spot Analysis and SaTScan identified an overlapping five-county cluster from 2006 to 2013. Poisson regression model, including county (inside vs. outside cluster), time (before vs. after 2006), and county*time indicators, was fit to evaluate the PRs of gastroschisis. The model did not confirm the presence of a spatiotemporal cluster, once it adjusted for county-level maternal characteristics (p = .549). CONCLUSION: Close monitoring of rates of gastroschisis is warranted to determine if the PRs of gastroschisis continue to decline in AR.

BackgroundUsing ICD-9 codes underestimates the prevalence of obesity in adults; however, the validity of these codes in studies of pregnancy-related outcomes is not known.ObjectivesTo compare classification of maternal obesity based on ICD-9 codes in hospital discharge records versus data from birth certificates in the same women, examine predictors of agreement, and assess how associations between obesity and two birth outcomes differ by source of weight data.MethodsThis population-based study included 2329145 California births between 2007 and 2012. We compared data on obesity from childbirth hospital discharge records (ICD-9 codes for obesity) and birth certificates (pre-pregnancy body mass index (BMI) calculated from weight and height) and identified predictors of agreement between the two sources. Logistic regression models assessed whether the two definitions of obesity resulted in different estimates of the associations of obesity with caesarean birth and large-for-gestational age.ResultsOverall, 464754 women (20.0%) had obesity based on their pre-pregnancy BMI while only 100002 (4.3%) had an obesity-related ICD-9 code. The sensitivity of ICD-9-based obesity was low at 16.2%; however, obesity codes were highly specific at 98.7%, with a negative predictive value of 82.5% and a positive predictive value of 75.2%. Among women with obesity identified by the birth certificate, those with pre-pregnancy and pregnancy-related complications (eg diabetes and hypertension) were more likely to have an obesity-related diagnosis in their delivery hospital discharge record. Using ICD-9 codes overestimated the association of obesity with caesarean birth and newborn large-for-gestational age.ConclusionsICD-9 codes in childbirth discharge records captured only one in five women with pre-pregnancy obesity. Sensitivity varied by maternal characteristics and conditions. This misclassification resulted in bias when examining the association of obesity and pregnancy-related outcomes.

Background Ministries of health, donors, and other decision‐makers are exploring how they can use mobile technologies to acquire accurate and timely statistics on births and deaths. These stakeholders have called for evidence‐based guidance on this topic. This review was carried out to support World Health Organization (WHO) recommendations on digital interventions for health system strengthening. Objectives Primary objective: To assess the effects of birth notification and death notification via a mobile device, compared to standard practice. Secondary objectives: To describe the range of strategies used to implement birth and death notification via mobile devices and identify factors influencing the implementation of birth and death notification via mobile devices. Search methods We searched CENTRAL, MEDLINE, Embase, the Global Health Library, and POPLINE (August 2, 2019). We searched two trial registries (August 2, 2019). We also searched Epistemonikos for related systematic reviews and potentially eligible primary studies (August 27, 2019). We conducted a grey literature search using mHealthevidence.org (August 15, 2017) and issued a call for papers through popular digital health communities of practice. Finally, we conducted citation searches of included studies in Web of Science and Google Scholar (May 15, 2020). We searched for studies published after 2000 in any language. Selection criteria For the primary objective, we included individual and cluster‐randomised trials; cross‐over and stepped‐wedge study designs; controlled before‐after studies, provided they have at least two intervention sites and two control sites; and interrupted time series studies. For the secondary objectives, we included any study design, either quantitative, qualitative, or descriptive, that aimed to describe current strategies for birth and death notification via mobile devices; or to explore factors that influence the implementation of these strategies, including studies of acceptability or feasibility. For the primary objective, we included studies that compared birth and death notification via mobile devices with standard practice. For the secondary objectives, we included studies of birth and death notification via mobile device as long as we could extract data relevant to our secondary objectives. We included studies of all cadres of healthcare providers, including lay health workers; administrative, managerial, and supervisory staff; focal individuals at the village or community level; children whose births were being notified and their parents/caregivers; and individuals whose deaths were being notified and their relatives/caregivers. Data collection and analysis For the primary objective, two authors independently screened all records, extracted data from the included studies and assessed risk of bias. For the analyses of the primary objective, we reported means and proportions, where appropriate. We used the Grading of Recommendations Assessment, Development and Evaluation (GRADE) approach to assess the certainty of the evidence and we prepared a 'Summary of Findings' table. For the secondary objectives, two authors screened all records, one author extracted data from the included studies and assessed methodological limitations using the WEIRD tool and a second author checked the data and assessments. We carried out a framework analysis using the Supporting the Use of Research Evidence (SURE) framework to identify themes in the data. We used the GRADE‐CERQual (Confidence in the Evidence from Reviews of Qualitative research) approach to assess our confidence in the evidence and we prepared a 'Summary of Qualitative Findings' table. Main results For the primary objective, we included one study, which used a controlled before‐after study design. The study was conducted in Lao People’s Democratic Republic and assessed the effect of using mobile devices for birth notification on outcomes related to coverage and timeliness of Hepatitis B vaccination. However, we are uncertain of the effect of this approach on these outcomes because the certainty of this evidence was assessed as very low. The included study did not assess resource use or unintended consequences. For the primary objective, we did not identify any studies using mobile devices for death notification. For the secondary objective, we included 21 studies. All studies were conducted in low‐ or middle‐income settings. They focussed on identification of births and deaths in rural, remote, or marginalised populations who are typically under‐represented in civil registration processes or traditionally seen as having poor access to health services. The review identified several factors that could influence the implementation of birth‐death notification via mobile device. These factors were tied to the health system, the person responsible for notifying, the community and families; and include: ‐ Geographic barriers that could prevent people’s access to birth‐death notification and post‐notification services ‐ Access to health workers and other notifiers with enough training, supervision, support, and incentives ‐ Monitoring systems that ensure the quality and timeliness of the birth and death data ‐ Legal frameworks that allow births and deaths to be notified by mobile device and by different types of notifiers ‐ Community awareness of the need to register births and deaths ‐ Socio‐cultural norms around birth and death ‐ Government commitment ‐ Cost to the system, to health workers and to families ‐ Access to electricity and network connectivity, and compatibility with existing systems ‐ Systems that protect data confidentiality We have low to moderate confidence in these findings. This was mainly because of concerns about methodological limitations and data adequacy. Authors' conclusions We need more, well‐designed studies of the effect of birth and death notification via mobile devices and on factors that may influence its implementation., Plain language summary Birth and death notification via mobile devices: a mixed methods review What is the aim of this review? In this Cochrane Review, we aimed to assess the effect of using mobile devices to notify births and deaths. We also aimed to describe how these mobile solutions are being used in practice and the factors that influence their use. We collected and analysed all relevant studies to answer these questions. Key messages We know very little about the effects of using mobile devices to notify births and deaths. Factors that can influence the implementation of this approach include factors tied to the health system and the notification system, the person responsible for notifying, the community, and the families involved. What was studied in the review? By registering notified births and deaths, governments can track the health of their population, identify needs and problems, and design better services. In many countries, births and deaths are not properly registered. Sometimes this is because government systems are poorly designed to facilitate registration, government workers do not have proper training, people live far from government services, or are not aware of the need to register births or deaths. In many cases, registration is affected by delays in or lack of birth or death notification. Governments are starting to use mobile devices such as mobile phones to reduce problems with birth or death notification. In some settings, members of the public, healthcare workers and others use mobile phones to notify a birth or death to the health system or to a central registration system. The main aim of our review was to assess what happens when people use mobile devices to notify births and deaths, compared to other systems or no systems at all. For instance, do they register more birth and deaths, and do they do this at the right time? And does this lead more babies and children to use or receive health services, such as immunisation? We also wanted to find out how people are using these mobile systems in practice and what influences their use. What are the main results of the review? We found 21 relevant studies. All of the studies were from low‐ or middle‐income countries. Most of these studies focussed on the notification of births and deaths in rural, remote, or marginalised populations who are often under‐represented in birth or death registration processes or have poor access to health services. Only one of the studies assessed the effect of using mobile devices for notification systems. This study focussed on birth notification. We did not find any studies that assessed the effect of using mobile devices for death notification. We are uncertain of the effect of this approach on the number of births and deaths that are properly notified because the certainty of this evidence was assessed as very low. The other studies had information about how people use the mobile device‐based birth and death notification systems in practice. These studies pointed to several factors that could influence the implementation of birth‐death notification via mobile devices. These factors were tied to the health system and the notification system, the person responsible for notifying, the community, and the families involved. They include: ‐ Geographic barriers that could prevent people’s access to birth‐death notification and post‐notification services ‐ Access to health workers and other notifiers with enough training, supervision, support, and incentives ‐ Monitoring systems that ensure the quality and timeliness of the birth and death data ‐ Legal frameworks that allow births and deaths to be notified by mobile device and by different types of notifiers ‐ Community awareness of the need to register births and deaths ‐ Socio‐cultural norms around birth and death ‐ Government commitment ‐ Cost to the system, to health workers and to families ‐ Access to electricity and network connectivity, and compatibility with existing systems ‐ Systems that protect data confidentiality We have low to moderate confidence in these findings. This was mainly because of the ways in which the studies were designed and small amounts of data. How up‐to‐date is this review? We searched for studies that had been published up to August 2, 2019.

Background Continuous electronic fetal monitoring (CEFM) is a standard of hospital care during the intrapartum period. We investigated its use on childbirth outcomes in low-risk pregnancies, and examined whether outcomes differed by gestational age within a term pregnancy. Methods A retrospective secondary data analysis using birth registry data from two diverse northeastern US states from 1992 to 2014. Chi-square test and the Fisher exact tests were used to examine associations between CEFM and childbirth outcomes. Multivariable Poisson regression models were used to estimate risk ratios of childbirth outcomes related to CEFM use, adjusting for potential confounders. Results Use of CEFM was independently associated with a 10% (State 1) and 40% (State 2) increased risk for primary cesarean delivery and an increased risk for assisted vaginal births (14% and 24%, respectively) after adjustment for confounders. CEFM use was not associated with reduced risk for infant mortality (neonatal mortality, 0-27 days, and post-neonatal mortality, 28-364 days) in term births (37-41 weeks' gestation). After stratifying term pregnancies into early term, full term, and late term, use of CEFM was associated with reduced risk for neonatal mortality in early-term births (37 0/7 weeks' to 38 6/7 weeks' gestation) in State 2 (RR 0.44 [95% CI 0.21-0.92]), but not in State 1. There was no association between CEFM use and infant mortality (neonatal and post-neonatal) in full-term or late-term births. Conclusions The study results do not support universal use of CEFM in pregnancies that are low-risk and at term.

BACKGROUND Rising cesarean rates call for studies on which subgroups of women contribute to the rising rates, both in countries with high and low rates. This study investigated the cesarean rates and contributing groups in Iceland using the Robson 10-group classification system. METHODS This study included all births in Iceland from 1997 to 2015, identified from the Icelandic Medical Birth Registry (81 839). The Robson distribution, cesarean rate, and contribution of each Robson group were analyzed for each year, and the distribution of other outcomes was calculated for each Robson group. RESULTS The overall cesarean rate in the population was 16.4%. Robson groups 1 (28.7%) and 3 (38.0%) (spontaneous term births) were the largest groups, and groups 2b (0.4%) and 4b (0.7%) (prelabor cesareans) were small. The cesarean rate in group 5 (prior cesarean) was 55.5%. Group 5 was the largest contributing group to the overall cesarean rate (31.2%), followed by groups 1 (17.1%) and 2a (11.0%). The size of groups 2a (RR 1.04 [95% CI 1.01-1.08]) and 4a (RR 1.04 [95% CI 1.01-1.07]) (induced labors) increased over time, whereas their cesarean rates were stable (group 2a: P = 0.08) or decreased (group 4a: RR 0.95 [95% CI 0.91-0.98]). CONCLUSIONS In comparison with countries with high cesarean rates, the prelabor cesarean groups (singleton term pregnancies) in Iceland were small, and in women with a previous cesarean, the cesarean rate was low. The size of the labor induction group increased, yet the cesarean rate in this group did not increase.

OBJECTIVE: Improving civil registration and vital statistics (CRVS) systems is essential to monitoring health objectives locally and globally. The barriers to birth and particularly death registration in low- and middle-income countries are however poorly understood.METHODS: We conducted a survey among women of reproductive age in Bissau, the capital of Guinea-Bissau. We asked women with a birth in the past two years whether their child had been registered and had obtained a birth certificate. We elicited the sources of information about birth registration and asked respondents to list their reasons for (not) registering a birth. If their child had died, we asked similar questions about death registration.RESULTS: Most women (86%) had received messages about birth registration, but few women whose child had died had heard about the need to register deaths (22%). The primary sources of information about birth registration were messages broadcast on the radio or displayed at health facilities. Information about death registration was primarily obtained through informal social networks. Only 16% of births, and 2% of deaths, had been registered. The main barriers to birth registration were administrative pre-requisites and paternal absence. The main reasons for not registering a death were lack of knowledge about death registration and lack of perceived benefits.CONCLUSION: Strengthening CRVS systems requires addressing the specific barriers preventing birth and death registration. In Bissau, interventions to improve knowledge about death registration are needed. Simplifying registration procedures, as well as providing additional incentives, might help improve the coverage of birth registration.

Background This study aims to examine hospital variation in both maternal and neonatal morbidities and identify institutional characteristics associated with hospital performance in a combined measure of maternal and neonatal outcomes. Methods Using the California Linked Birth File containing data from birth certificate and hospital discharge records, we identified 1 322 713 term births delivered at 248 hospitals during 2010-2012. For each hospital, a risk-standardized rate of severe maternal morbidities and a risk-standardized rate of severe newborn morbidities were calculated after adjusting for patient clinical risk factors. Hospitals were ranked based on combined information on their maternal and newborn morbidity rates. Results Risk-standardized severe maternal and severe newborn morbidity rates varied substantially across hospitals (10th to 90th percentile range = 67.5-148.2 and 141.8-508.0 per 10 000 term births, respectively), although there was no significant association between the two (P = 0.15). Government hospitals (non-Federal) were more likely than other hospitals to be in worse rank quartiles (P value for trend = 0.004), whereas larger volume was associated with better rank among hospitals in the first three quartiles (P = 0.004). The most prevalent morbidities that differed progressively across hospital rank quartiles were severe hemorrhage, disseminated intravascular coagulation, and heart failure during procedure/surgery for mothers, and severe infection, respiratory complication, and shock/resuscitation for neonates. Conclusions Hospitals with low maternal morbidity rates may not have low neonatal morbidity rates and vice versa, highlighting the importance of assessing joint maternal-newborn outcomes in order to fully characterize a hospital's obstetrical performance. Hospitals with smaller volume and government ownership tend to have less desirable outcomes and warrant additional attention in future quality improvement efforts.

BACKGROUND In many low- to middle-income countries (LMIC) assisted vaginal birth rates have fallen, while caesarean section (CS) rates have increased, with potentially deleterious consequences for maternal and perinatal mortality. AIMS To review birth mode and perinatal mortality in a large LMIC hospital with strict labour management protocols and expertise in vacuum extraction. MATERIALS AND METHODS We conducted a retrospective observational study at Port Moresby General Hospital in Papua New Guinea. Birth registers from 1977 to 2015 (39 years) were reviewed. Overall and modified (fresh stillbirths and early neonatal deaths ≥500 g) perinatal mortality rates (PMRs) were calculated by birthweight/birth mode. RESULTS There were 365 056 births (5215 in 1977; 14 927 in 2015), of which 14 179 (3.9%) were vacuum extractions, 609 (0.2%) forceps births and 14 747 (4.4%) CS (increase from 2% to 5%). The failure rate of vacuum extraction was 2.5% (range 0.5-5.4%). Symphysiotomy was employed for 184 births. From 1989 to 2015, the modified mean PMR for babies ≥2500 g was 8.1/1000 births (range 5.6-12.1; 6.9 in 2015), 9.1/1000 for babies ≥1500 g (7.3-14.8; 9.1 in 2015) and 7.5/1000 (0-21.7; 9.0 in 2015) for vacuum extractions (98% were ≥2500 g). The overall PMR for these years was 29.7/1000 births. CONCLUSIONS In an LMIC with rapidly increasing birth numbers a comparatively low PMR can be achieved while maintaining low CS rates. This may be in part accomplished through strict use of second-stage protocols, perinatal audit, and supportive training that promotes judicious and proficient use of vacuum extraction and CS.

BACKGROUND Zika virus has recently emerged as a novel cause of microcephaly. CDC has asked states to rapidly ascertain and report cases of Zika-linked birth defects, including microcephaly. Massachusetts added head circumference to its birth certificate (BC) in 2011. The accuracy of head circumference measurements from state vital records data has not been reported. METHODS We sought to assess the accuracy of Massachusetts BC head circumference measurements by comparing them to measurements for 2,217 infants born during 2012-2013 captured in the Massachusetts Birth Defects Monitoring Program (BDMP) data system. BDMP contains information abstracted directly from infant medical records and served as the true head circumference value (i.e., gold standard) for analysis. We calculated the proportion of head circumference measurements in agreement between the BC and BDMP data. We assigned growth chart head circumference percentile categories to each BC and BDMP measurement, and calculated the sensitivity and specificity of BC-based categories to predict BDMP-based categories. RESULTS No difference was found in head circumference measurements between the two sources in 77.9% (n = 1,727) of study infants. The sensitivity of BC-based head circumference percentile categories ranged from 85.6% (

Background Classifying preterm birth as spontaneous or indicated is critical both for clinical care and research, yet the accuracy of classification based on different data sources is unclear. We examined the accuracy of preterm birth classification as spontaneous or indicated based on birth certificate data. Methods This is a retrospective cohort study of 123 birth certificates from preterm births in Missouri. Correct classification of spontaneous or indicated preterm birth subtype was based on multi-provider (RN, MFM Fellow, MFM attending) consensus after full medical record review. A categorisation algorithm based on clinical data available in the birth certificate was designed a priori and classification was performed by a single investigator according to the algorithm. Accuracy of birth certificate classification as spontaneous or indicated was compared to the consensus classification. Errors in misclassification were explored. Results Classification based on birth certificates was correct for 66% of preterm births. Most errors in classification by birth certificate occurred in classifying a birth as spontaneous when it was in fact indicated. The vast majority of errors occurred when preterm rupture of membranes (≥12 h) was checked on the birth certificate causing classification as spontaneous when there was a maternal or fetal indication for delivery. Conclusions Birth certificate classification overestimated spontaneous preterm birth and underestimated indicated preterm birth compared to classification performed from medical record review. Revisions to birth certificate clinical data would allow more accurate population level surveillance of preterm birth subtypes.

SummaryObjective Seizures are a common manifestation of neurologic dysfunction in neonates and carry a high risk for mortality and adverse long-term outcomes. U.S. birth certificates are a potentially valuable source for studying the epidemiology of neonatal seizures. However, the quality of the data is understudied. Methods We reviewed all U.S. birth records from 2003 to 2013 to describe the following: (1) rates of missing data, (2) evidence of underreporting, and (3) effect of the 2003 revision of the birth certificate form. We evaluated missingness by state, year, demographic, infant health, and medical care factors using bivariate analyses. To measure potential underreporting, we compared estimates to a published reference (0.95 per 1,000 term births). We developed criteria for data plausibility, and reported which states met these criteria. Results Of 22,834,395 live term births (≥36 weeks of gestation) recorded using the revised form from 2005 to 2015, there were 5,875 with neonatal seizures, suggesting an incidence of 0.26 per 1,000 term births, one fourth of the expected incidence. Although the overall degree of missing seizure data was low (0.5%), missingness varied significantly by state, year, demographic, infant health, and medical care factors. After the 2003 birth certificate form revision, missing data and evidence of potential underreporting increased. Nine states met criteria for plausibility. Significance The value of U.S. birth certificate data for neonatal seizure epidemiology is limited by biased missingness, evidence suggestive of underreporting, and changes in reporting subsequent to the 2003 revision. There are plausible data from nine states, which merit investigation for further research.

Objective To improve on existing methods to infer race/ethnicity in health care data through an analysis of birth records from Connecticut. Data source A total of 162 467 Connecticut birth records from 2009 to 2013. Study design We developed a logistic model to predict race/ethnicity using data from US Census and patient-level information. Model performance was tested and compared to previous studies. Five performance measures were used for comparison. Principal findings Our full model correctly classifies 81 percent of subjects and shows improvement over extant methods. We achieved substantially improved sensitivity in predicting black race. Conclusions Predictive models using Census information and patients' demographic characteristics can be used to accurately populate race/ethnicity information in health care databases, enhancing opportunities to investigate and address disparities in access to, utilization of, and outcomes of care.

Background Gestational age estimation by last menstrual period (LMP) vs. ultrasound (or best obstetric estimate in the US) may result in discrepant classification of preterm vs. term birth. We investigated whether such discrepancies are associated with adverse infant outcomes. Methods We studied singleton livebirths in the Medical Birth Registries of Norway, Sweden and Finland and US live birth certificates from 1999 to the most recent year available. Risk ratios (RR) with 95% confidence intervals (CI) by discordant and concordant gestational age estimation for infant, neonatal and post-neonatal mortality, Apgar score

Objective: Evidence of identity, particularly a birth certificate, is essential to access many rights. However, the births of many Aboriginal Australians are not registered when they are infants. We examined factors related to birth registration among Western Australian children born to Aboriginal mothers. Methods: All births to Aboriginal mothers in the Midwives Notification System in Western Australia (WA) from 1980 to 2010 were linked to birth registrations. Associations between registration and maternal and child characteristics were examined for children aged under 16 years in 2012. Results: Among 49,694 births between 1980 and 2010, 18% of those aged under 16 years had unregistered births, compared to 3% of those aged 16–32 years. Unregistered births were most strongly associated with young maternal age at first birth (adjusted odds ratio [AOR] 5.22; 95%CI 3.07–8.86; for 16 years or younger vs 30 years or older, among non-smokers), remoteness (AOR 2.17; 95%CI 1.87–2.52; very remote vs major cities), mothers whose own birth was unregistered (AOR 3.00; 95%CI 1.78–5.07) and no private hospital insurance (AOR 0.19; 95%CI 0.11–0.31; insured vs uninsured). Conclusions: Unregistered births are common among WA Aboriginal children, particularly in disadvantaged families. Implications: Assistance before discharge from hospital may increase birth registrations.

Objective Questions have been raised about potential risks of prenatal exposure to mercury from amalgam fillings during pregnancy. The aim of this study was to assess possible associations between exposure to amalgam fillings in pregnant women participating in a large cohort study and adverse pregnancy outcome. Methods In the Norwegian Mother and Child Cohort Study (MoBa), a valid information about the number of teeth with amalgam fillings and dental treatment, including new amalgam fillings placed or removed during pregnancy, was available from 69 474 pregnancies. The information was obtained from two questionnaires sent to the women at 17 and 30 weeks of pregnancy, and the data were linked to the Medical Birth Registry of Norway. Logistic regression was used to estimate the odds ratio (OR) and 95% confidence intervals (95% CI) as a measure of association between pregnancy outcome and prenatal exposure to amalgam fillings. Results Logistic regression models, including mothers′ age, education, BMI, parity, smoking during pregnancy, and alcohol consumption during pregnancy revealed no significant associations between the number of teeth with amalgam fillings and early preterm delivery, late preterm delivery, low birthweight, malformation or stillbirth. Conclusions We found no evidence for serious perinatal consequences of maternal exposure to amalgam fillings during pregnancy.

Objective To investigate associations between mode of delivery and subsequent reproductive outcomes. Design Cohort study. Population Women with term singleton live births from 1987 to 2009. Setting Denmark, birth registration data. Methods Women with a first singleton delivery after 37 weeks were followed until the end of 2010, from a first birth to include subsequent live births. We used Cox's proportional hazards model stratified by parity to compare the likelihood for subsequent delivery according to mode of delivery at first and later births, estimating maternal age effects and lag time to next delivery. Main outcome measure Likelihood of a subsequent live-born child by previous delivery mode. Results We identified 642 052 women with a first delivery. Compared with women with a non-instrumental vaginal delivery, delivering a child by elective cesarean section implied a 23% (95% CI 0.76–0.787) decreased likelihood for subsequent delivery. Emergency cesarean section meant 16% fewer (95% CI 0.84–0.85), and vaginal instrumental delivery 4% fewer subsequent deliveries (95% CI 0.95–0.96). Hazard ratios were largely unchanged after controlling for parity and year of birth. Small age-trends were seen, with hazard ratios affected by maternal age at birth. Delivery mode at first birth affected marginally the time lag until next birth. Conclusions Fecundity, measured as likelihood of a successive live-born child, varied with mode of delivery at the first and also subsequent births. A first or later delivery by cesarean section implied decreased likelihood of subsequent delivery compared with women with a first vaginal birth.

Objective To study the incidence, maternal characteristics and outcome of unplanned out-of-institution births (= unplanned births) in Norway. Design Register-based cross-sectional study. Population All births in Norway (n = 892 137) from 1999 to 2013 with gestational age ≥22 weeks. Methods Analysis of data from the Medical Birth Registry of Norway from 1999 to 2013. Unplanned births (n = 6062) were compared with all other births (reference group). Results The annual incidence rate of unplanned births was 6.8/1000 births and remained stable during the period of study. Young multiparous women residing in remote municipalities were at the highest risk of experiencing unplanned births. The unplanned birth group had higher perinatal mortality rate for the period, 11.4/1000 compared with 4.9/1000 for the reference group (incidence rate ratio 2.31, 95% confidence interval 1.82–2.93, p

Objective: Birth records and hospital admission records are valuable for research on maternal smoking, but individually are known to under-estimate smokers. This study investigated the extent to which combining data from these records enhances the identification of pregnant smokers, and whether this affects research findings such as estimates of maternal smoking prevalence and risk of adverse pregnancy outcomes associated with smoking. Methods: A total of 846,039 birth records in New South Wales, Australia, (2001–2010) were linked to hospital admission records (delivery and antenatal). Algorithm 1 combined data from birth and delivery admission records, whereas algorithm 2 combined data from birth record, delivery and antenatal admission records. Associations between smoking and placental abruption, preterm birth, stillbirth, and low birthweight were assessed using multivariable logistic regression. Results: Algorithm 1 identified 127,612 smokers (smoking prevalence 15.1%), which was a 9.6% and 54.6% increase over the unenhanced identification from birth records alone (prevalence 13.8%), and delivery admission records alone (prevalence 9.8%), respectively. Algorithm 2 identified a further 2,408 smokers from antenatal admission records. The enhancement varied by maternal socio-demographic characteristics (age, marital status, country of birth, socioeconomic status); obstetric factors (multi-fetal pregnancy, diabetes, hypertension); and maternity hospital. Enhanced and unenhanced identification methods yielded similar odds ratios for placental abruption, preterm birth, stillbirth and low birthweight. Conclusions: Use of linked data improved the identification of pregnant smokers. Studies relying on a single data source should adjust for the under-ascertainment of smokers among certain obstetric populations.

To describe the epidemiology of cerebral palsy (CP) in western Sweden.A population-based study covering 94 466 live births in the area in 2003-2006. Birth characteristics and neuroimaging findings were recorded, prevalence was calculated and aetiology was analysed.CP was found in 206 children, including postneonatal cases, corresponding to a crude prevalence of 2.18 per 1000 live births. The gestational age-specific prevalence for28 gestational weeks was 71.4 per 1000 live births, while it was 39.6 for 28-31 weeks, 6.4 for 32-36 weeks and 1.41 per 1000 for36 weeks. Hemiplegia accounted for 44%, diplegia for 29% and tetraplegia for 6%, while 16% had dyskinetic CP and 5% had ataxia. Neuroimaging was available in 95% of the children. This showed maldevelopment in 13%, white matter lesions in 36%, cortical/subcortical lesions in 23% and basal ganglia lesions in 14%. The aetiology was considered to be prenatal in 36% and perinatal/neonatal in 46% and remained unclassified in 18%.The overall prevalence of CP in western Sweden was stable. However, the distribution of CP types changed and the term hemiplegia increased significantly. Among children with CP born extremely preterm, the percentage born before 26 weeks of gestation had increased.

Introduction Although out-of-hospital births are still relatively rare in the United States, it is important to monitor trends in these births, as they can affect patterns of facility usage, clinician training, and resource allocation, as well as health care costs. Trends and characteristics of home and birth center births are analyzed to more completely profile contemporary out-of-hospital births in the United States. Methods National birth certificate data were used to examine a recent increase in out-of-hospital births. Results After a gradual decline from 1990 to 2004, the number of out-of-hospital births increased from 35,578 in 2004 to 47,028 in 2010. In 2010, 1 in 85 US infants (1.18%) was born outside a hospital; about two-thirds of these were born at home, and most of the rest were born in birth centers. The proportion of home births increased by 41%, from 0.56% in 2004 to 0.79% in 2010, with 10% of that increase occurring in the last year. The proportion of birth center births increased by 43%, from 0.23% in 2004 to 0.33% in 2010, with 14% of the increase in the last year. About 90% of the total increase in out-of hospital births from 2004 to 2010 was a result of increases among non-Hispanic white women, and 1 in 57 births to non-Hispanic white women (1.75%) in 2010 was an out-of-hospital birth. Most home and birth center births were attended by midwives. Discussion Home and birth center births in the United States are increasing, and the rate of out-of-hospital births is now at the highest level since 1978. There has been a decline in the risk profile of out-of-hospital births, with a smaller proportion of out-of-hospital births in 2010 than in 2004 occurring to adolescents and unmarried women and fewer preterm, low-birth-weight, and multiple births.

Objective To determine the risk factors associated with having a very low birth weight (VLBW) infant as a follow-up to the first phase of a Perinatal Periods of Risk approach. Design and Sample Retrospective cohort analysis of birth certificates. Population-based sample of 53,427 birth certificates for the city under study during the years 1999–2006. Measures The relationship of selected maternal characteristics as predictors of VLBW using multivariate logistic regression analysis. Results The maternal characteristics associated with VLBW were as follows: no prenatal care (OR = 4.04), inadequate weight gain (OR = 3.97), Black, non-Hispanic race (OR = 1.50), less than 20 years old (OR = 1.42) and more than 35 years old (OR = 1.43). After analyzing age and race/ethnicity together, Black non-Hispanic women less than 20 years of age (OR = 2.70) or over 35 years of age (OR = 2.45) still had an increased odds for having a VLBW infant whereas Black non-Hispanic women between the ages of 20 and 35 did not. Conclusions The findings of this study suggest educating women on the importance of preconception care, prenatal care, and adequate pregnancy weight gain to reduce the odds of having a VLBW infant.

Racial and ethnic differences and disparities in health and health care are issues of growing concern that need to be properly identified and classified to be adequately studied and addressed. Yet health plans often do not readily have race/ethnicity information at their disposal for a large proportion of their members (High-Value Health Care Project 2010; Weissman and Hasnain-Wynia 2011). This may be due to limitations, such as infrequent contact with members, added costs of developing the systems to collect such information, perceptions of regulatory prohibitions against collecting race/ethnicity, and discomfort or fear among members in providing it (America’s Health Insurance Plans and Robert Wood Johnson Foundation’s 2004; Institute of Medicine 2009; Escarce et al. 2011; Weissman and Hasnain-Wynia 2011; Gazmararian et al. 2012). Although many health plans are beginning to collect race/ethnicity data more systematically, it is likely to take a substantial amount of time before this information is complete (Escarce et al. 2011). While self-report is considered the gold standard (Institute of Medicine 2009; Gazmararian et al. 2012), several techniques have been developed to indirectly estimate race/ethnicity where it is unavailable, which would allow for monitoring and evaluation of its role in health care and health research in the meantime (High-Value Health Care Project 2010). This approach has been encouraged by the United States (US) government’s Agency for Healthcare Research and Quality and the Institute of Medicine (Institute of Medicine 2009). One such recent method is the Bayesian Improved Surname Geocoding method (BISG) developed by Rand Corporation, which has been demonstrated to improve upon previous techniques (Elliott et al. 2009). The algorithm, described in detail by Elliott et al. (2008, 2009), utilizes a Bayesian approach to combine racial/ethnic data from last names and geographic units. While the use of indirect race/ethnicity estimations can enhance research efforts, an assessment of the quality of these data in a variety of contexts is needed before implementation within each setting. Some validation studies comparing the BISG algorithm estimates to self-reported race/ethnicity have been performed by its developers (Elliott et al. 2008, 2009; Elliott 2009; High-Value Health Care Project 2010) as well as by health plans beginning to utilize it. However, these studies have included limited additional stratification by pertinent personal characteristics; in particular, analysis of the BISG algorithm’s performance by age is lacking. Moreover, while the intended use of the BISG algorithm is to directly apply the predicted probabilities and not to classify individuals into a single race/ethnicity, it is important to understand how the algorithm performs for individual assignment of race/ethnicity in instances where the use of probability values is not feasible. For example, when attempting to identify individuals of a certain race/ethnicity for participation in a targeted program or research study, it will be necessary to decide an appropriate probability level that would warrant contacting individuals for recruitment. Yet analyses to determine optimal cutoffs on which to base such determinations have not been reported. In this study, we therefore assessed the validity of BISG algorithm estimations for the “off-label” use of assigning race/ethnicity within a regional managed care organization (MCO) population and examined how well the algorithm performed by gender and age.

Objective To describe risk factors for recurrent preterm birth (PTB) in the second and third birth. Design Historical cohort study. Setting Utah, USA. Population Women who had their first three singleton live births in Utah between 1989 and 2007 and a preterm first or second birth were included. Methods Maternally linked birth records were used. Multivariable-adjusted risk ratios were calculated for recurrent PTB. Results were stratified by spontaneous and indicated PTB and by pattern of birth outcomes. Main outcome measures Risk ratios and 95% confidence intervals for risk factors for recurrent PTB. Results Among women with PTB in their first or second live birth, recurrent PTB occurred in 21% of second live births (n = 1011/4805) and 22% of third live births (n = 1872/8468). Risk factors for recurrence included short inter-pregnancy interval, underweight prepregnancy body mass index, pre-existing maternal medical conditions, history of PTB at 28–32 weeks of gestation (versus 33–36 weeks), the presence of a fetal anomaly, and young maternal age. Risk factors for spontaneous, but not indicated PTB included young maternal age and less than appropriate gestational weight gain. Risk factors also varied in women experiencing a first versus second recurrence in their third birth. Conclusions Risk factors may vary by the clinical subtype of the most recent PTB and the pattern of term and preterm outcomes across births 1–3; some of the risk factors identified in this study may be modifiable through interventions targeted at women in the inter-conception period.

BACKGROUND Maternal self-report is the most common method for assessment of past cigarette exposure to assess birth defect risk. This study compared maternal smoking prior to and during pregnancy based on self-reports obtained from the medical records abstracted for the Utah Birth Defect Network (UBDN), the birth certificate, and the computer-assisted telephone interview (CATI) in the National Birth Defects Prevention Study (NBDPS). The study also investigated how the different sources for maternal smoking data affect estimates in an empirical study. METHODS A total of 1774 case and 618 control mothers who had participated in the NBDPS and whose live born infants were delivered between January 1, 2003, and December 31, 2007, were included in this study. Among the case mothers, we compared data from all three sources, whereas for control mothers only two data sources were available for comparison (i.e., birth certificate and CATI). RESULTS Smoking prevalence was highest in the CATI. Compared to the CATI, data from the UBDN had a higher sensitivity (61.3%) and better agreement (kappa = 0.63) than birth certificates (51.8%; kappa = 0.56). Adjusted odds ratios for all and specific birth defects (i.e., holoprosencephaly, hydrocephalus, anophthalmia/microphthalmia, anotia/microtia, total anomalous pulmonary venous return/partial anomalous pulmonary venous return [TAPVR/PAPVR], heterotaxy, and gastroschisis) were different between the birth certificate and CATI. The change in the effect estimates between the two sources ranged from 19% to 56%. CONCLUSIONS Based on our findings, maternal smoking exposure from interview data was shown to be of higher quality with less misclassification compared to data obtained from medical records or birth certificates. Birth Defects Research (Part A) 2012. © 2012 Wiley Periodicals, Inc.

SUMMARY This is a first effort to quantify the contribution of different factors in explaining racial difference in low birthweight rate (LBW). Mother's health, child characteristics, prenatal care, socioeconomic status (SES), and the socioeconomic and healthcare environment of mother's community are important inputs into the birthweight production function, and a vast literature has delved into obtaining causal estimates of their effect on infant health. What is unknown is how much of the racial gap in LBW is explained by all these inputs together. We apply a nonlinear extension of the Oaxaca–Blinder method proposed by Fairlie to decompose this gap into the portion explained by differences in observed characteristics and the portion that remains unexplained. Data are obtained from several sources in order to capture as many observables as possible, although the primary data source is the Natality Detail Files. Results show that of the 6.8 percentage point racial gap in LBW, only 0.9–1.9 points are explained by white–black differences in endowments across those measures, and of those endowments, most of the gap in LBW is explained by the differences in SES. The unexplained difference is attributed to racial differences in the returns to or the marginal product of investments in infant health. Copyright © 2011 John Wiley & Sons, Ltd.

Objective To compare prenatal care coverage between adolescent (early and late) and adult pregnant women in Iquitos, Peru. Methods A random sample of 4467 birth records was collected. Multivariate analyses were performed to compare prenatal care coverage in all adolescent (10–14 years, 15–19 years) and adult (≥ 20 years) age groups and then for primiparous women only. Results The mean number of visits was 5.0 for adolescents aged 10–14 years, 6.1 for adolescents aged 15–19 years, and 6.2 for women aged 20 years or older. For primiparous women, the means were 5.1, 6.2, and 6.8, for the respective age groups. Both the proportion attending and the number of prenatal visits were significantly lower in primiparous adolescents aged 10–14 years than in primiparous women aged 20 years or older (aOR 0.25; 95% CI, 0.10–0.62 and aRR 0.83; 95% CI, 0.74–0.94, respectively). Conclusion All women attended prenatal care more frequently than the WHO's recommended 4 visits; however, early adolescents attended significantly less often than late adolescents or adult women. Further study of this inequality is warranted to adequately inform local health services.

BACKGROUND: Smoking in pregnancy increases the risk for many different adverse pregnancy outcomes, including birth defects. Gastroschisis, a birth defect most commonly associated with young maternal age has been associated with smoking, but findings are inconsistent. We assessed whether smoking increases the risk for gastroschisis using population-based data from Utah. METHODS: Gastroschisis cases (n = 189) were identified from the Utah Birth Defect Network and all live births without birth defects (n = 423,499) occurring in Utah from January 1, 1997 through December 31, 2005 served as controls. Exposure data were derived from birth certificates and fetal death certificates and, for terminated pregnancies, the Utah Birth Defect Network. RESULTS: Women who smoked during the first trimester of pregnancy had an increased risk of gastroschisis (OR 1.6; 95% CI: 1.1, 2.3) after adjusting for maternal age and preconception BMI. Discordance between birth certificate data and data from structured interviews increased exposure prevalence from 16.9 to 22.2% for case mothers and 7.4 to 13.2% for control mothers. Accounting for this misclassification, the crude OR decreased by 24%, 1.9 (1.3, 2.7). CONCLUSIONS: Though first trimester cigarette smoking was reported on birth certificates by more mothers of gastroschisis cases than controls, adjustment for confounders (maternal age and preconception BMI) and smoking misclassification suggests the association is weak. Despite a decrease in smoking prevalence among all women of childbearing years in Utah between 1997 and 2005, the prevalence of gastroschisis has not followed a similar trend. Birth Defects Research (Part A), 2008. © 2008 Wiley-Liss, Inc.

Most midwives are aware of the need to collect clinical practice data and of its usefulness in supporting the care they provide, which contributes to healthy outcomes for mothers and babies. For the individual midwife, there is more than one easily accessible, standardized data collection instrument from which to choose. However, despite these choices, in an American College of Nurse-Midwives (ACNM) Division of Research (DOR) survey on midwifery clinical data collection (N = 263), the majority of member respondents (n = 135; 51%) reported using a self-designed data collection tool, and more than one-third did not know of the ACNM Nurse-Midwifery Clinical Data Sets (NMCDS). On a larger scale, the midwifery profession is also in need of an organized and consistent approach to data collection for the purpose of capturing midwifery practice and outcomes in order to provide data to support legislation, practice, and policy changes. However, the profession currently lacks a single common midwifery practice database. In order to facilitate data aggregation that captures a larger view of midwifery practice at the local, regional, and national levels, it is imperative that all midwives collect relevant data that are uniform and standardized, and that the midwifery professional organizations move forward with the development of a common electronic database. This article describes currently available data collection tools as well as their best uses, applications to practice, and future directions.

We analysed a transgenerational linked birth file to investigate the relationship between maternal birthweight and infant birthweight-specific mortality risk for white and African American infants. Birth records of 267,303 infants born between 1989 and 1991 were linked to records of their mothers, born between 1956 and 1976, and to their own death certificates for those dying in the first year. The means, standard deviations and z-scores were calculated for each race- and generation-specific birthweight distribution. Investigators then analysed the mortality of very small infants (birthweight at least two standard deviations below their mean) for three maternal birthweight categories. Over half of the infant deaths involved births with weights more than two standard deviations below the relevant population mean birthweight (comprising 4.2% of white and 6.9% of African American births respectively). African American infants experienced higher mortality rates at all levels of standardised birthweight, from z-scores of -3 to +3. The relative risk of mortality associated with very small infant size was less for infants delivered to smaller birthweight mothers when compared with those whose mothers were average sized or large at birth. This differential effect was confined to neonatal deaths and was more prominent in the white subpopulation.

In this study we examined the relationship between diagnoses of isolated intrapartum fever or chorioamnionitis and the risk of encephalopathy in term newborns. We conducted a population-based, case-control study in Washington State using 1994 to 2002 linked data from the Washington State Birth Registry and the Comprehensive Hospital Abstract Reporting System (CHARS). We identified 1060 singleton, term newborns (602 males, 458 females) with International Classification of Diseases (ICD-9) diagnoses consistent with encephalopathy, and 5330 unaffected control newborns (2756 males, 2574 females). Intrapartum fever was defined by a diagnosis of intrapartum temperature of >38 degrees C in the birth registry or CHARS databases. Chorioamnionitis was defined using ICD-9 diagnoses recorded in CHARS. We identified 2.2 cases of encephalopathy per 1000 births. Isolated intrapartum fever was associated with a 3.1-fold (95% confidence interval [CI] 2.3-4.2) increased risk of newborn encephalopathy. Chorioamnionitis was associated with a 5.4-fold (95% CI 3.6-7.8) increased risk of encephalopathy. We found that isolated intrapartum fever and chorioamnionitis were independently associated with an increased risk of encephalopathy in term infants. Our data also indicate that there is a spectrum of risk for encephalopathy in term infants exposed to intrapartum fever. Infants born to women with signs of chorioamnionitis other than isolated intrapartum fever may be at higher risk of encephalopathy than those exposed only to isolated intrapartum fever.

Accurate estimation of gestational age early in pregnancy is paramount for obstetric care decisions and for determining fetal growth and other conditions that may necessitate timing the iatrogenic intervention or delivery. We sought to examine temporal changes in the distributions of two measures of gestational age, namely, those based on menstrual dating and a clinical estimate. We further sought to evaluate relative comparisons and variability in indices of perinatal outcomes. We utilised the Natality data files in the US, 1990-2002 comprising women that delivered a singleton livebirth between 22 and 44 weeks gestation (n = 42 689 603). Changes were shown in the distributions of gestational age based on menstrual vs. clinical estimate between 1990 and 2002, as well as changes in the proportions of preterm (37,32 and28 weeks) and post-term (or=42 weeks) birth, and small- (SGA;10th percentile) and large-for-gestational-age (LGA; birthweight90th percentile) births. While the absolute rates of preterm birth37 weeks, SGA and LGA births were lower based on the clinical estimate of gestational age relative to that based on menstrual dating, the increases in preterm birth rate between 1990 and 2002 were fairly similar between the two measures of gestational dating. However, the decline in post-term births was larger, based on the clinical estimate (-73.8%), than on the menstrual estimate (-36.6%) between 1990 and 2002. While the clinical estimate of gestational age appears to provide a reasonably good approximation to the menstrual estimate, disregarding the clinical estimate of gestational age may ignore the advantages of gestational age assessment in modern obstetrics.

Although early ultrasound (20 weeks' gestation) systematically underestimates the gestational age of smaller fetuses by approximately 1-2 days, this bias is relatively small compared with the large error introduced by last menstrual period (LMP) estimates of gestation, as evidenced by the number of implausible birthweight-for-gestational age. To characterise this misclassification, we compared gestational age estimates based on LMP from California birth certificates with those based on early ultrasound from a California linked Statewide Expanded Alpha-fetoprotein Screening Program (XAFP). The final sample comprised 165 908 women. Birthweight distributions were plotted by gestational age; sensitivity and positive predictive value for preterm rates according to LMP were calculated using ultrasound as the 'gold standard'. For gestational ages 20-27 and 28-31 weeks, the LMP-based birthweight distributions were bimodal, whereas the ultrasound-based distributions were unimodal, but had long right tails. At 32-36 weeks, the LMP distribution was wider, flatter, and shifted to the right, compared with the ultrasound distribution. LMP vs. ultrasound estimates were, respectively, 8.7% vs. 7.9% preterm (37 weeks), 81.2% vs. 91.0% term (37-41 weeks), and 10.1% vs. 1.1% post-term (or=42 weeks). The sensitivity of the LMP-based preterm birth estimate was 64.3%, and the positive predictive value was 58.7%. Overall, 17.2% of the records had estimates with an absolute difference of14 days. The groups most likely to have inconsistent gestational age estimates included African American and Hispanic women, younger and less-educated women, and those who entered prenatal care after the second month of pregnancy. In conclusion, we found substantial misclassification of LMP-based gestational age. The 2003 revised US Standard Certificate of Live Birth includes a new gestational age item, the obstetric estimate. It will be important to assess whether this estimate addresses the problems presented by LMP-based gestational age.

Summary Correspondence: Michelle Pearl, SequoiaFoundation c/o GeneticDisease Screening Program,California Department ofPublic Health, 850 Marina BayParkway, Rm. F175, Mail Stop8200, Richmond, CA 94804,USA.E-mail:michelle.pearl@cdph.ca.gov Pearl M, Wier ML, Kharrazi M. Assessing the quality of last menstrual period dateon California birth records. Paediatric and Perinatal Epidemiology 2007; 21 (Suppl. 2):50–61.Birth certificate last menstrual period (LMP) date is widely used to estimate gestationalage in the US. While data quality concerns have been raised, no large population-basedstudyhasisolateddataqualityissuesbycomparingbirthrecordLMP(BirthLMP)withreliable LMP dates from another source. We assessed LMP data quality in 2002 Cali-fornia singleton livebirth records ( n = 515 381) and in a subset of records with linkedprenatally collected LMP from California’s statewide Prenatal Expanded Alpha-fetoprotein Screening Program (XAFP) ( n = 105 936). Missing or incomplete LMP dataaffected 13% of birth records; 17% of those had complete LMP within XAFP records.Data quality indicators supported XAFP LMP as more accurate than Birth LMP, witha lower prevalence of digit preference, post-term delivery, out-of-range gestational ageestimates and implausible birthweight-for-gestational age. The bimodal birthweightdistribution evident at 20–31 weeks’ gestation based on Birth LMP was nearly absentwith XAFP LMP-based gestational age. Approximately 32% of the second birthweightmode was explained by apparent clerical errors in Birth LMP month. Digit preferenceerrors, particularly day 1, were associated with gestational age overestimation. Pretermdelivery rates were higher according to Birth (7.6%) vs. XAFP LMP (7.2%). One-fifth ofobservedpretermandoverhalfofobservedpost-termbirthsusingBirthLMPwerenottrue cases; 15% of true preterm cases were missed. African American or Hispanic, lesseducated, and publicly or uninsured women were most likely to be misclassified andhave large LMP date discrepancies attributable to clerical or digit preference error.Theimplementationofarevisedbirthcertificateisanopportunityfortargetedtrainingand data entry checks that could substantially improve LMP accuracy on birthrecords.

Gestational age on the birth certificate is the most common source of population-based gestational age data that informs public health policy and practice in the US. Last menstrual period is one of the oldest methods of gestational age estimation and has been on the US Standard Certificate of Live Birth since 1968. The 'clinical estimate of gestation', added to the standard certificate in 1989 to address missing or erroneous last menstrual period data, was replaced by the 'obstetric estimate of gestation' on the 2003 revision, which specifically precludes neonatal assessments. We discuss the strengths and weaknesses of these measures, potential research implications and challenges accompanying the transition to the obstetric estimate.

Estimates of the gestational age of the newborn based on US Birth Certificate data are extensively used to monitor trends in infant and maternal health and to improve our understanding of adverse pregnancy outcome. Two measures of gestational age, the 'date of the last normal menses' (LMP) and the 'clinical estimate of gestation' (CE), have been available from birth certificate data since 1989. Reporting irregularities with the LMP-based measure are well-documented, and important questions remain regarding the derivation of the CE. Changes in perinatal medicine and in vital statistics reporting in recent years may have importantly altered gestational age data based on vital statistics. This study describes how gestational age measures are collected and edited in US national vital statistics, and examines changes in the reporting of these measures by race and Hispanic origin between 1990 and 2002. Data are drawn from the National Center for Health Statistics' restricted use US birth files for 1990-2002. Bivariable statistics are used. The percentage of records with missing LMP dates declined markedly over the study period, overall, and for each racial/Hispanic origin group studied. A marked shift in the distribution of the CE of gestational age was also observed, suggesting changes both in the true distribution of age at birth, and in the derivation of this measure. Agreement between the LMP-based and CE estimates increased over the study period, especially among preterm births. However, a high proportion of LMP dates continue to be missing or invalid and the derivation of the CE is still uncertain. In sum, although the reporting of gestational age measures in vital statistics appears to have improved between 1990 and 2002, substantial concerns with both the LMP-based and the CE persist. Efforts to identify approaches to further improve upon the quality of these data are needed.

Nurse-midwifery has accomplished remarkable clinical, policy, and political achievements using specially-collected data. Today, midwifery practice data can be found in existing administrative data systems: birth registration, hospital data depositories, and claims files. Issues in finding midwifery as practice and profession in these data systems are discussed. Improving the integrity of data that reveal midwives as caregivers should be a priority.

Epidemiological surveillance and research on birth defects require accurate diagnosis and adequate registration. In this regard, the performance of national birth registries is not well described.We linked clinical data from all 3,616 cleft cases treated in Norway between 1967 and 1998 with data from the Medical Birth Registry of Norway, and calculated the proportion of clinically verified cases reported to the Registry, stratified by severity.The cleft type most completely ascertained was cleft lip and palate (CLP), of which 94% were reported. Ascertainment was less complete for cleft lip alone (83% recorded), and cleft palate only (CPO) (57% recorded). For each of the three types of clefts, completeness of reporting depended on severity of the cleft. For example, 71% of cases with severe CPO were reported, while only 11% of cases with mild CPO were reported.Ascertainment was strongly related to cleft type and severity. To the degree that severity of birth defects may be related to their cause, these patterns of registration have implications for surveillance of birth defects as well as the conduct of etiologic studies. The large proportion of cleft palate cases unrecorded at birth suggests that clinical examination of the newborn palate is often inadequate.

BACKGROUND: Although birth defects are a leading cause of death in infancy and early childhood, the proportion of all deaths to children with clinically diagnosed birth defects is not well documented. The study is intended to measure the proportion of all deaths to infants and children under age 10 occurring to children with birth defects and how and why this proportion differs from the proportion of deaths due to an underlying cause of congenital anomalies using standard mortality statistics. METHODS: A linked file of Michigan livebirths and deaths was combined with data from a comprehensive multisource birth defects registry of Michigan livebirths born during the years 1992 through 2000. The data were analyzed to determine the mortality rate for infants and children with birth defects and for children with no reported birth defect. Mortality risk ratios were calculated. The underlying causes of death for children with birth defects were also categorized and compared to cause- specific mortality rates for the general population. RESULTS: Congenital anomalies were the underlying cause of death for 17.8% of all infant deaths while infants with birth defects were 33.7% of all infant deaths in the study. Almost half of all Michigan deaths to children aged 1 to 2 were within the birth defects registry, though only 15.0% had an underlying cause of death of a congenital anomaly based upon standard mortality statistics. The mortality experience among children with birth defects was significantly higher than other children throughout the first 9 years of life, ranging from 4.6 for 5 year olds to 12.8 for children 1 to 2. Mortality risk ratios examined by cause of death for infants with birth defects were highest for other endocrine (28.1), other CNS (28.1), and heart (21.9) conditions. For children 1 through 9, the highest differential risk was seen for other perinatal conditions (39.0), other endocrine (29.7), other CNS (24.5), and heart (21.4). CONCLUSIONS: Childhood mortality analyses that incorporate birth defects registry data provide a more comprehensive picture of the full burden of birth defects on mortality in infant and children and can provide an effective mechanism for monitoring the survival and mortality risks of children with selected birth defects on a population basis. Birth Defects Research (Part A) 79:792–797, 2007. © 2007 Wiley-Liss, Inc.

The aim was to assess the neonatal morbidity of alternative modes of delivery using economic data. Two groups of neonatal morbidity data were extracted according to mode of delivery from inpatient claims on National Health Insurance in Taiwan: uncomplicated vaginal and caesarean section deliveries. Outcome variables included number of infants treated as inpatients duration of hospital stay and hospital charge during both the first month and the first year of age. Uncomplicated caesarean section delivery of term infants carried a significant increase either in the first month of life or during the first year after birth in both duration of hospital stay and hospital discharge when compared with uncomplicated vaginal delivery. Neonatal jaundice accounted for the greatest morbidity among term newborns irrespective of delivery mode followed by infectious gastrointestinal and respiratory morbidity. Logistic regression analysis demonstrated that uncomplicated vaginal delivery was associated with skin diseases. Uncomplicated caesarean section delivery was associated with infectious disease and disease of the respiratory digestive and circulatory systems. The findings of this study may provide further information for clinicians and would be an important consideration when advising pregnant women on the preferred route of delivery. (authors)

Background: Nationally and in New Jersey, the cesarean delivery rate has been increasing steadily for nearly a decade, and especially since 1999. The purpose of this study was to describe recent trends in cesarean section delivery in New Jersey. Methods: Data on delivery method, medical indications and patient characteristics were extracted from electronic birth certificate files. Results: Cesarean section deliveries increased as a proportion of live births by 6 percent annually. Growth was roughly uniform across Robson’s clinical classification. Repeat cesareans contributed only proportionately to the overall trend. The greatest acceleration was observed for procedures without trial of labor, and in medical situations where cesarean delivery had been relatively rare. Conclusions: Medical indications recorded on the birth certificate explained little of the rapid growth in utilization of cesarean delivery, since trends were comparable in most categories we examined. A sustained autonomous shift in practice patterns, patient preferences, or both seems the most likely driver of the overall trend. (BIRTH 33:3 September 2006)

Background. In this study, 2 main hypotheses have been put forward to explain the variation in childhood intelligence or school performance by season of birth. In the first hypothesis, it is suggested that it is due to school policy concerning school entry, whereas the second suggests that a seasonally patterned exposure such as temperature, maternal nutrition, or infection during critical periods of brain development have a lasting effect on intelligence. Aims. To determine whether childhood performance on tests of different domains of intelligence is patterned by season of birth and to examine possible mechanisms for any associations. Sample. 12,150 individuals born in Aberdeen, Scotland between 1950 and 1956. Methods. Birth cohort study in which the variation in different domains of childhood intelligence measured at ages 7, 9, and 11 by season of birth were examined. Results. Reading ability at age 9 and arithmetic ability at age 11 varied by season of birth, with lowest scores among those born in autumn or early winter (September–December) and highest scores among those born in later winter or spring (February–April); p = .002 for joint sine-cosine functions for reading ability at age 9 and p = .05 for sine-cosine function for arithmetic ability at age 11. The child's perception and understanding of pictorial differences at age 7, verbal reasoning at 11, and English language ability at 11 did not vary by season of birth. Age at starting primary school and age relative to class peers were both associated with the different measurements of childhood intelligence and both attenuated the association between month of birth and reading ability at age 9 and arithmetic ability at age 11 towards the null. Both adjusted and unadjusted differences in reading ability at age 9 and arithmetic ability at age 11 between those born from September to December compared with other times of the year were less than 0.1 of a standard deviation of the test scores. Ambient temperature around the time of conception, during gestation, and around the time of birth did not affect intelligence. Conclusion. Any variation in mean childhood intelligence by season of birth is weak and largely explained by age at school entry and age relative to class peers.

Vital statistics documents for the United States have been revised approximately decennially throughout the past century. In this commentary we review the contents of the 2003 revision of the national standard certificate of live birth, focusing on changes from the certificates in use nationally since 1989, and identifying strengths and weaknesses of key data elements. Additional federal-state partnership funding is imperative to support the transition to the new standard certificates, and to ensure a focus on data quality at the state and national levels.

Summary Linkage of routinely collected health data collections is increasingly being used to investigate maternal and infant morbidity and mortality. Such data have the advantage of being population based and readily available. However, in using such data it is important to understand the data linkage process, the proportions of unmatched records and the characteristics of these records so that potential bias can be recognised. This article describes the differences in characteristics of matched and unmatched mothers’ and babies’ records generated in the linkage of birth records with hospital discharge data and explores some of the reasons for these differences. The study population included over 250 000 women and their babies discharged from hospital following delivery in New South Wales, Australia between 1 January 2000 and 31 December 2002. Hospital discharge and birth data were linked using probabilistic linkage methods for both mothers and babies. Matching rates were 98.5% and 99.0% for maternal birth and hospital discharge records, respectively, and 98.8% and 99.4% for baby records. Unmatched maternal records had higher proportions of Australian-born women, private hospital births and stillbirths compared with matched records. Unmatched baby records had higher proportions of low-birthweight babies, preterm births and in-hospital deaths than matched records. With the possible exception of stillbirths, these differences are unlikely to cause important bias in studies relying on matched records only. Our results suggest studies using linked data should generally examine and report on the characteristics of unmatched records, and recognise them as a potential source of bias.

Although the neonatal mortality rate (NMR) in Bangladesh remained steady between 1995-99 and 1999-2003 (41-42 deaths per 1,000 live births), evidence from the management information system (MIS) of a large nongovernmental organization (NGO) program indicates that the NMR declined by about 50 percent between 1996 and 2002 in the area served. This study aims to validate the recording of neonatal deaths among the cohort of children registered as born in 2003 and to assess the evidence of a decline in the NMR. It also measures the coverage of reproductive health outreach services, focusing on 12 of the 27 NGOs that have provided services in the same areas since 1996. Field-workers' registers, verbal autopsy reports, and immunization records were checked to confirm infants' survival. Interviews were conducted with 142 mothers of children who died within 28 days postpartum and with a random sample of 109 women with registered stillbirths. Out of 11,253 registered live births in 2003, 210 neonatal deaths were found, compared with 194 deaths that were reported in the MIS for 2003. The corrected NMR was 19 deaths per 1,000 live births, and it was in the range of 15-29 deaths per 1,000 live births in 11 of the NGO areas. Because underreporting of neonatal deaths was probably higher in 1996 when the MIS-reported NMR was 39 deaths per 1,000 live births, the decline in the NMR is likely to have been genuine.

Background: Healthy breastfeeding practice in the United States depends decisively on high rates of initiation at the delivery hospital. We sought to estimate the component of hospital variation in rates of exclusive breastfeeding at discharge that was dependent on demographic composition. Isolating that component can help to illuminate the potential independent contribution of hospital policies, practices, and staff behaviors. Methods: Electronic birth certificate data in New Jersey from 1996 to 2001 (n = 545,837) were used to measure variations in hospital-level rates of breastfeeding initiation. The method of infant feeding within 24 hours before hospital discharge was reported as exclusive breastfeeding, formula feeding, combination feedings, other methods, and unknown. Rates of exclusive breastfeeding by hospital were standardized to remove sociodemographic differences in hospitals’ service populations that influence initiation rates. Results: Sociodemographic variables predicted about 60 percent of the variation in hospital-specific rates of exclusive breastfeeding at discharge. Hospitals that were designated intensive or regional perinatal centers, delivered higher volumes of infants, and served more breastfeeding-prone populations were only slightly more likely to have higher adjusted rates compared with other hospitals; considerable unexplained variation remained. Conclusions: Standardized exclusive breastfeeding rates pointed to the contribution of population demographics to breastfeeding initiation, and other contributions, including hospital practices, are also important. To protect, promote, and support breastfeeding, a more detailed evidence base on hospital policies and practices should be developed, and hospitals should review their policies and practices in light of documented best breastfeeding practice. (BIRTH 32:2 June 2005)