MTHFR) C677T genotype, with the birth weight of offspring.We used Mendelian randomization analysis to investigate the causal relationship between maternal homocysteine level, as represented by maternal methylenetetrahydrofolate reductase . Birth weight decreased from 120 to 130 grams as maternal homocysteine level increased, while controlling for confounding factors; however, the association was of marginal significance (P = 0.06).Our results suggest an adverse relationship between maternal homocysteine level and birth weight. A reduction in homocysteine levels might positively affect birth outcomes. In addition to these risk factors, homocysteine concentration was reported to be positively associated with the risk of pregnancy complications2 and low birth weight.3 In general, homocysteine concentration remains low during pregnancy.2 However, abnormal homocysteine levels in the intrauterine environment could induce adverse birth outcomes such as pregnancy loss,4,5 placental abruption,5 and intrauterine growth retardation.3The extent of intrauterine growth can determine lifelong health and the health of future generations. During recent decades, a number of studies have confirmed Barker’s hypothesis regarding the fetal origin of adult disease.MTHFR) genotype.2MTHFR is a critical enzyme required for the synthesis of 5-methyltetrahydrofolate, an active form of folic acid in blood. It is involved in converting homocysteine to methionine,6 and a known polymorphism of MTHFR is a transition of C to T at nucleotide 677 of the MTHFR gene, which is a cause of reduced MTHFR activity7 and hyperhomocysteinemia.8 Moreover, the Hordaland Homocysteine Study reported a trend toward restricted intrauterine growth as the number of T alleles increased.9 In addition, previous studies reported that women homozygous for C677T in MTHFR had an increased risk of preeclampsia and placental vasculopathy.2,3 Although previous studies showed that MTHFR C677T mutation could lead to hyperhomocysteinemia and retardation of intrauterine growth, there is no direct evidence that variation in maternal homocysteine due to MTHFR genotype affects offspring growth in utero. One approach to investigate causal inference, Mendelian randomization, is an application of instrumental variable (IV) analysis that uses genetic instruments, due to the principal of independent assortment of an allele during meiosis. The use of Mendelian randomization in an observational study was introduced by Katan.10 In the present study, we use maternal genotype as an unconfounded and unbiased instrument for homocysteine level in maternal blood. The application of IV analysis using maternal genetic variants allows us to estimate intergenerational effects.Homocysteine metabolism is affected by many factors, including age, gender, current diet, and methylenetetrahydrofolate reductase .n = 720). For all participants, venous blood was collected from the antecubital vein during the morning after an overnight fast, and all participants responded to structured questionnaires. We excluded women who had twin births (n = 30), gestational diabetes (n = 23), hypertension, or chronic renal disease (n = 13), those with no data on homocysteine or folate level due to inadequate blood sampling (n = 92), and those with incomplete questionnaires (n = 89). Therefore, a total of 473 newborns were analyzed for this study . There was no difference between participants and excluded subjects in the frequencies of the CC, CT, and TT genotypes . This study was approved by the Institutional Review Board of Ewha Womans University Hospital.We recruited women between 24 and 28 weeks’ gestation who visited Ewha Womans University Hospital in Seoul, Republic of Korea for prenatal care during the period from August 2001 to December 2003. All participants gave their informed consent for study participation. Our analysis of deliveries during the period from October 2001 to February 2004  showed that 2481 women gave birth in Ewha Womans University Hospital. About 30% of them were included in this study (MTHFR C677T genotypes were identified by polymerase chain reaction (PCR) amplification and digestion with Hinf1. The PCR product resulted in synthesis of 198 bp fragments, which contain a C to T substitution at nucleotide 677 of the MTHFR gene. The fragments were cut into 175 and 23 bp fragments by the derived T allele and were visualized on 2.5% polyacrylamide gel with ethidium bromide staining after electrophoresis.Extraction of DNA from fasting venous blood was done using a QIAmp DNA blood kit  according to the protocol recommend by the manufacturer, and samples were stored at −20°C until analysis. 18 column. Folate was measured by a radioimmunoassay kit .High-performance liquid chromatography was used to measure homocysteine. Briefly, we mixed 100 µL of plasma with 10 µL of 10% tri-n-butylphosphine and stored the sample for 30 min at 4°C. Deproteinization was performed by adding 100 µL of 10% trichloroacetic acid, followed by centrifugation for 5 minutes at 3000 g. The supernatant was mixed with 250 µL of 0.125 M borate buffer and 100 µL of ammonium 7-fluorobenzo-2-oxa-1,3-diazole-4-sulfonate acid. After incubation for 1 hour at 60°C, the mixture was filtered with a 0.45-µm syringe filter . Fluorescence detection was performed  using an Xterra RPBirth weight and length were measured in the delivery room by trained nurses using digital scales. Gestational age was calculated based on the maternal report of the date of last menstrual period and ultrasound measurement performed by an obstetrician.T alleles (in the codominant model) and as a continuous variable (in the additive model). We used a general linear model and the χ2 test to assess associations of confounding variables with the different MTHFR genotypes.Data analysis was conducted using the statistical package STATA (version 11.0). In descriptive analysis, continuous variables were presented as mean with SD and categorical variables as frequencies. Genotype was analyzed both as a categorical variable based on number of MTHFR C677T genotype was used as an IV to estimate the association between intrauterine homocysteine environment and the birth weight of offspring. Gestational age, the sex of the offspring, maternal age, and folate concentration were analyzed as confounding variables. We used the Durbin form of the Durbin–Wu–Hausman statistic to compare estimates of the effects of homocysteine on birth weight obtained from ordinary least-squares linear regression and IV analysis.11Mendelian randomization analysis was done using the “ivreg” command in STATA. Two-stage least-squares estimates were used to fit the IV analysis. Maternal CC, CT, and TT in the included  and excluded  subjects were similar and were in Hardy–Weinberg equilibrium (P > 0.05). The mean homocysteine concentration of the subjects was 5.76 µmol/L and did not significantly differ from that of excluded subjects  , and mean birth weight was 3299 grams. The genotype frequencies of .2%, and MTHFR C677T genotype and showed a dose–response association for each additional T allele: namely, CC (5.47 µmol/L), CT (5.75 µmol/L), and TT (6.48 µmol/L)  after controlling for gestational age and sex. In particular, adjusted mean birth weight for the TT allele was lower than that for the CC allele . However, homocysteine level was not significantly associated with birth weight, even after adjustment for confounding variables (data not shown).Birth weight significantly differed in relation to P = 0.06). Ordinary least-squares regression analysis showed a clear distinction in the direction of the associations, and the results of IV analysis and the Hausman test were statistically significant (P < 0.05). This suggests that, as compared with the ordinary least-squares method, IV analysis is more suitable for analyzing the relationship of maternal homocysteine and birth weight (data not shown).Table MTHFR C677T genotype and homocysteine level. There was a dose–response association between homocysteine and homozygosity for MTHFR C677T (TT), ie, homocysteine level significantly increased with the number of T alleles (P for trend <0.01). Furthermore, the findings of our advanced statistical approach suggest that maternal homocysteine concentration during pregnancy has an adverse effect on the birth weight of offspring, although the result was of marginal significance (P = 0.06), perhaps due to limited statistical power or the small sample size. To our knowledge, no other study has used an advanced statistical approach to investigate the association of intrauterine homocysteine exposure, as determined by maternal MTHFR genotype, with offspring birth weight.We found a strong association between MTHFR genotype to estimate the causal effect of variation in maternal homocysteine. The characteristics of Mendelian randomization permit genetic variants to be investigated in an observational study. Because such studies have properties similar to those of intention-to-treat analyses in randomized controlled trials,12 the problems of observational studies, eg, reverse causality and bias due to confounding factors, are avoided.13 In addition, genetic variants have larger effects on intermediate features than on the disease itself.14 The MTHFR C677T polymorphism is already established as an instrument: a study using Mendelian randomization reported that homocysteine level, as determined by MTHFR C677T gene variant, was associated with an increased risk for stroke.15As mentioned above, the use of maternal gene polymorphism as an instrumental variable in our analysis allowed us to use 3 A causal role for homocysteine in the pathogenesis of metabolic disorders has been suggested in hypertension, congenital heart disease, and cardiovascular disease.3,8,15 In addition, many previous studies have reported that homocysteine was associated with adverse effects on pregnancy and birth outcomes2,5,16; however, opinions differ in relation to causality. Regarding the genetic effect on homocysteine concentration, a transition of C to T at nucleotide 677 of the MTHFR gene was shown to cause hyperhomocysteinemia, which could inhibit MTHFR enzyme activity. Moreover, this inhibition of MTHFR enzyme activity might be amplified by folate deficiency.2 As in previous studies, we found that homocysteine negatively correlated with folate concentration in blood  and that homocysteine level was inversely associated with offspring birth weight, even after controlling for folate concentration in blood and other potential confounders. Studies reported that babies with MTHFR mutation were small and had low birth weights, due to growth restriction caused by MTHFR dysfunction.17,18 Moreover, several studies reported that variation in MTHFR C677T genotype resulted in different pregnancy outcomes. Newborns homozygous for C677T have approximately double the risk of adverse birth outcomes .19 Therefore, future studies should examine whether co-occurrence of MTHFR C677T in mother and fetus increases the risk of undesirable pregnancy outcomes.Homocysteine is normally found in blood, and it has been suggested that increased levels impair endothelial vasomotor function and facilitate platelet production, thereby damaging arteries and blood flow.20 For that reason, a number of studies have used birth weight as a main outcome. In our previous study, we found that birth weights of offspring from mothers with hyperhomocysteinemia (≥15 µmol/L) were significant lower than those of offspring from mothers with normal homocysteine concentrations.21 However, that study did not adjust for relevant confounding factors and did not use an adequate statistical approach, such as multilevel modeling for path construction. Folic acid supplementation has been proposed as a treatment of hyperhomocysteinemia. A randomized control trial observed that oral folic supplementation significantly reduced plasma homocysteine level.22 Other studies reported that consumption of folic acid attenuated the risk of preterm birth in women with pre-eclampsia.16,23 However, lowering homocysteine level does not decrease disease risk. Hence, future studies should examine the true effect of folic acid supplementation on adverse pregnancy outcomes.The importance of birth weight to lifelong health has been a consistent theme during the last few decades. Birth weight is a good surrogate marker of intrauterine growth retardation caused by diverse factors. Low birth weight increases the risk of chronic diseases and adversely affects the health of subsequent generations.This study had several limitations. First, in a genetic epidemiology study the results of Mendelian randomization may be confounded by genetic variants in linkage disequilibrium with the variant of interest. In addition, unmeasured and residual confounding factors, such as nutritional intake during pregnancy, may have affected the results. A number of factors in the model, including homocysteine and maternal age, are associated with miscarriage and adverse birth outcomes. Because only women who successfully completed pregnancy were included in the present study, there is a possibility of selection bias. Last, due to lack of statistical power, generalizability of the results is problematic—similar previous studies enrolled more than 1000 people. Despite the relative small size of the present study, we obtained valuable results from IV analysis. Future studies should investigate the effect of homocysteine on birth outcomes by considering the genotypes of many people.24 The present advanced statistical method also allows for inference of causality. In addition, by considering a genetic polymorphism as an instrumental variable, it can assess unmeasurable effects. In conclusion, our results provide more evidence of an adverse relationship between maternal homocysteine level and birth outcomes and suggest that reduction of homocysteine levels during pregnancy could positively affect birth outcomes.A strength of the present study is that it used an innovative study methodology that differed from the designs of previous studies, which were mostly case-control studies.


Congenital laryngeal cyst is a rare cause of airway obstruction that may require urgent diagnosis and treatment. We report a case of a neonate having history of polyhydramnios and severe respiratory distress at birth. A laryngeal cyst detected during intubation. The outcome of laryngoscopic treatment of the cyst was favorable.  Benign congenital laryngeal cysts are rare entities with an estimated incidence of <2 per 100,000 live births [1]. They can cause severe respiratory distress immediately after birth [2]. The etiology is not clear. With improvement in fetal imaging modalities, laryngeal cysts can be diagnosed antenatally in the second trimester [3], which helps in planning immediate perinatal management. Undiagnosed cases of laryngeal cyst are associated with high mortality (40%), thus underscoring the importance of antenatal diagnosis. We present a case of newborn with laryngeal cyst detected during intubation who received successful treatment with microlaryngoscopic excision and cauterization.A 2900 g, term male newborn, with history of maternal polyhydramnios was admitted to the NICU because of severe cyanosis and respiratory insufficiency that developed immediately after birth. During intubation, a large laryngeal mass was noted. Laryngotracheobronchoscopy (LTB) under general anesthesia revealed a cystic mass in the left supraglottic region of the larynx . The cyst was punctured and the mucoid fluid drained by suction to prevent aspiration. In the next few days after extubation, progressive inspiratory stridor developed. A second LTB showed a large recurrent cyst. Microlaryngoscopic excision of the prominent part of the cyst was performed . Pathologic examination was consistent with a laryngeal cyst. For one week, the neonate could not be extubated. Consequently, peroral corticosteroid therapy and inhalation therapy with salbutamol and ipratropium bromide were administered. After a few days, the neonate was extubated and discharged well on day 30 of life. Congenital laryngeal cysts are rare but may be fatal [4]. Apart from acute life threatening airway obstruction, presenting symptoms include stridor, hoarse cry, apnea, cyanosis, chest retractions, feeding problems and failure to thrive. Severe antenatal obstruction may lead to polyhydramnios and pulmonary hypoplasia. Compression of the trachea, cervical vessels and hypoglossal nerve have also been reported [1]. Antenatally, the diagnosis can be suspected on ultrasonography, which can show either the cyst or polyhydramnios [5]. In our case, polyhydramnios was detected prenatally but laryngeal cyst was not detected. Fortunately, it was detected during intubation performed for respiratory insufficiency. Polyhydramnios complicates 4% of pregnancies [6]. Polyhydramnios is idiopathic in approximately 35% to 65% of cases. However, with advancement in technology, the cause can be detected in most of the cases. In case of polyhydramnios, the pharynx and neck should be investigated thoroughly in order to identify lesions that could cause neonatal ventilation problems. About 18% of cervical congenital abnormalities can lead to polyhydramnios [7]. Congenital laryngeal cysts are rare, but are a possible cause of early neonatal respiratory distress. Due to their size and position, they can cause total or subtotal airway obstruction with consequent respiratory insufficiency and require emergency endotracheal intubation as happened in the index case [8]. Surgical procedures that are proposed for the treatment of congenital laryngeal cysts vary: While smaller lesions can be excised, marsupialised during endoscopy, larger or recurrent cysts may require a lateral cervical approach. In our patient, laryngoscopic drainage, excision of the prominent part and cauterization were performed. Microlaryngoscopic excision in our case assisted with endoscopic cauterization was safe and effective, with minimal morbidity. To conclude, in a neonate with respiratory insufficiency and a prenatal history of polyhydramnios, one should always consider the possibility of congenital laryngeal cysts and be alert that a careful observation during intubation can lead to diagnosis and prompt life saving management.Source of Support: NilConflict of Interest: None declared


GHRfl) or exon 3 deleted (GHRd3). An association of these isoforms has been found with small for gestational age (SGA) infants but does not influence adult height. The role of this polymorphism in the birth size spectrum in the general population is unclear.Human growth hormone receptor (GHR) transcripts have two isoforms, full-length  and the Moore normal pregnancy cohort (n = 274). GHR variants, wild-type (fl) and deleted for exon 3 (d3) were analysed using multiplex PCR.Pregnant women from white European families were recruited by the University College London Foetal Growth Study (fl/fl (36%) and overrepresentation of d3/d3 (14%) genotypes in the SGA infants within the cohorts . Fl/fl was overrepresented in large for gestational age (LGA) infants . There was a significant association of infants GHR isoforms with placental weight (P < 0·001) and birth weight standard deviation scores (P = 0·04) with the fl/fl genotype associated with a larger placental and birth weight. In multiple regression analysis, the GHR isoform type, maternal booking weight and parity influenced placental weight . The GHR isoform type was not related to antenatal anthropometric measurements or growth in infancy.There was a significant underrepresentation of infants wild-type GHR isoforms are associated with placental and birth weight.These data suggest that the  GHR transcripts exist in two major isoforms; retention (GHRfl) or exclusion of exon 3 (GHRd3). GHR exon 3 skipping results from a homologous recombination of two retroviral sequences flanking this exon that mimic an alternative splicing event.GHR exon 3 insertion/deletion as common polymorphism with these isoforms specific to individuals4The growth hormone receptor (GHR) gene (MIM: *600946) is located on chromosome 5 5p13.1-p12) and consists of nine coding exons.3.1-p12 aGHR with insertion/deletion of an exon could explain some of the variability seen in the antenatal and postnatal growth and response to recombinant human GH (r-hGH) treatment in children with growth disorders.During infancy and childhood pituitary, growth hormone (hGH-N) and insulin-like growth 1 (IGF-1) influence growth by virtue of hGH-N binding to the GHR. During pregnancy, hGH-N expression in the mother is suppressed; and placental growth hormone (hGH-V), a GH variant expressed by the placenta, becomes the predominant GH in the mother.et al.GHR exon 3 polymorphism on response to treatment with human GH in children with idiopathic short stature (ISS) and those born small for gestational age (SGA). However, no difference in the frequency of this polymorphism was found in short children compared to control adults of normal height, suggesting that this polymorphism was not primarily related to the aetiology of adult height.GHR polymorphism on growth rate. Subsequently, a number of studies of the polymorphism frequencies in various populations of children with growth disorders have yielded inconsistent resultsGHRd3 stimulates growth velocity by an additional effect of 0·5 cm during the first year of treatment with r-hGH.7Dos Santos GHR exon 3 polymorphism on antenatal and early postnatal growth with small sample sizes limiting interpretation.GHR exon 3 isoform on antenatal growth, birth size, placental weight and postnatal growth up to 3 years of age in two large prospectively recruited white European cohorts.At present, little is known about the role of n = 1650) and the Moore cohort (n = 310). Ethical approval was obtained from the University College London Hospitals and Queen Charlotte Hospital for Women and Hammersmith Hospital Ethics Committees. Written informed consent was obtained from all the participants on entrance into the study and then again for the participation of the infant in the study after birth. Suitable DNA samples for this study were available in 1048 infants, 774 from UCL-FGS and 274 from Moore cohort. Inclusion criteria in both cohorts were white European families presenting for the first prenatal visit before 20 weeks of pregnancy and ultrasound examination demonstrating a structurally normal single foetus. Pregnancies with antenatal complications and adverse foetal findings were excluded from the study.Subjects for this study were recruited from two separate white European cohorts, the University College London Hospital Fetal Growth Study (UCL-FGS) (n = 524) was used for the study. Data collection was standardized, and all data were collected prospectively. Antenatal growth measurements were performed by ultrasonography by a single skilled physician with expertise in foetal medicine. At birth, trimmed placental weight  was recorded. Anthropometric measures were undertaken at birth by a single observer. Weight was measured using electronic self-calibrating scales , length by Infantometer  and head circumference using a flexible metal tape. Three separate measurements were taken and the mean recorded. All measurements at birth were corrected for gestational age and sex by converting them into standard deviation scores (SDS).From UCL-FGS antenatal ultrasonographic growth data, anthropometric measurement at birth and growth data until 3 years of age .12n = 449) and from placenta (n = 325) where suitable cord blood sample was not available. In the Moore cohort, genomic DNA samples were obtained from placental tissue (n = 274). Placental samples were dissected near the umbilical cord insertion point, snapped frozen in liquid nitrogen and stored at −80 °C until use for foetal DNA preparation. DNA was extracted from the maternal blood samples and stored for analyses of maternal genotype.In the UCL-FGS cohort, genomic DNA was extracted from cord blood . Amplification products were analysed by electrophoresis on a 1% agarose gel stained with ethidium bromide. GHR fl allele was detected as a single band corresponding to 935 bp and the GHR-d3 allele corresponding to 532 bp, respectively. All samples were coded, and the operator was blinded to anthropometric data.For genotyping of the t-test was used to compare the GHR genotyped group (n = 774) and total cohort (n = 1674) mean values to ensure there was no selection bias in those samples used for genotype analysis.All data were assessed for normal distribution by exploring the data for skewness and kurtosis and the Shaprio–Wilks estimates. Infants were defined as SGA and large for gestational age (LGA) if birth weights were below 10th centile or above 90th centile, respectively, for gestational age. Postnatal catch-up or catch-down growth was defined as an increase or decrease in weight SDS of 0·6 over the first year of life.GHR genotypes. One-way analysis of variance (anova), with the Tukey’s honest significant differences (HSD) post hoc test, was used to determine differences between the mean anthropometric measures in the GHR exon 3 genotypes. Multiple stepwise linear regression was used to determine factors influencing placental weight and birth weight SDS.Chi-square tests were used to compare frequency distributions of the A sample size of 1048 was estimated to have a power of 90% to calculate a 50 g difference in placental weight and 100 g difference in birth weight at 5% level of significance between SGA and LGA groups.n = 774) and those remaining participants who did not (data not shown). All the data had normal distribution and hence log transformation was not required.Both cohorts consisted of a total of 1048 singletons with equal number of men and women. Maternal and infant cohort data shown in GHR exon 3 genotype was as follows: fl/fl 48%, fl/d3 45% and d3/d3 7%. The genotype distribution was in Hardy–Weinberg equilibrium. Parity and cigarette smoking were distributed similarly across the GHR genotype .The frequency distribution of the GHR genotype was not associated with parameters of antenatal growth  and birth weight SDS . In infant’s placenta, weight and birth weight SDS was higher in the GHR fl/fl genotype group and lower in the GHR d3/d3 genotype. GHR genotype was not significantly associated with birth length SDS  or head circumference SDS .In the UCL-FGS, l growth . The effd3/d3 (14%) and lower proportion of fl/fl (36%) in the SGA group at birth . In the LGA category the fl/fl genotype was overrepresented (60%) and fl/d3 and d3/d3 genotypes were underrepresented  (There was a greater proportion of  df = 2) .GHR genotype was found with postnatal anthropometric data up to 3 years of age (data not shown). GHR genotype was also not associated with catch-up  or catch-down  growth in the first year of life.In the UCL-FGS, no association of R2 35%), GHR genotype, booking weight, parity and gestational age of delivery were associated with placental weight while maternal age and cigarette smoking did not influence placental weight (GHR genotype (data not shown).In multiple regression analyses . Further, the effect does not seem to be maintained into postnatal life as no long-term effect on size could be determined nor on the catch-up and catch-down growth process.In a cohort of white European pregnant women and their offspring, we found a significant positive association between PGH gene in the foetus is associated with marked foetal growth restriction and a small placentaGHRfl/fl group in our study was heavier than that of the GHRd3/d3 group.Adequate placentation and placental growth are essential for foetal growth. There is a strong correlation between placental weight and foetal growth with most low birth infants having a small placenta.GHR genotype, while birth weight SDS is not suggesting that GHR genotype effect on birth weight could be mediated through placental weight.Our data would suggest that the size of the effect on birth weight is less than that upon the placenta but given the interaction of placental weight and birth size it is not easy to dissect this further. In the multiple linear regression model, placental weight is influenced by et al.GHR exon 3 deleted genotype on foetal growth velocity in SGA infants and increased postnatal growth in both SGA and appropriate for gestational age (AGA) infants. Sorensen et al.GHR d3/d3 genotype similar to our study. Unlike other studies, we could not demonstrate any effect on prenatal weight up to 32 weeks of gestational age and postnatal growth. This probably reflects the fact that our cohort represents a continuum across the birth size spectrum in the general population and does not represent polarized size class groups. In addition, the larger sample size in our study reduces the likelihood of false-positive findings.Jensen GHR locus on height or metabolic parameters.GHR variation does not significantly influence growth or the SNP that tags GHR exon 3 in the GWAS does not recapitulate all the genomic variability at the locus, including the d3, other copy number variations, or rare variants.21Of note, large genome wide association studies (GWAS) have not found any influence of GHR exon 3 polymorphism genotype is associated with placental weight and birth weight with carriers of GHRfl/fl allele having a heavier placenta and higher birth weight. Placental weight could be potentially influenced by an autocrine/paracrine action of hGH-V or other products of the chorio-somatomammotropic gene on the GHR polymorphism present in the placenta.In conclusion, we found that the


An efficient spectral element (SE) with electric potential degrees of freedom (DOF) is proposed to investigate the static electromechanical responses of a piezoelectric bimorph for its actuator and sensor functions. A sublayer model based on the piecewise linear approximation for the electric potential is used to describe the nonlinear distribution of electric potential through the thickness of the piezoelectric layers. An equivalent single layer (ESL) model based on first-order shear deformation theory (FSDT) is used to describe the displacement field. The Legendre orthogonal polynomials of order 5 are used in the element interpolation functions. The validity and the capability of the present SE model for investigation of global and local responses of the piezoelectric bimorph are confirmed by comparing the present solutions with those obtained from coupled 3-D finite element (FE) analysis. It is shown that, without introducing any higher-order electric potential assumptions, the current method can accurately describe the distribution of the electric potential across the thickness even for a rather thick bimorph. It is revealed that the effect of electric potential is significant when the bimorph is used as sensor while the effect is insignificant when the bimorph is used as actuator, and therefore, the present study may provide a better understanding of the nonlinear induced electric potential for bimorph sensor and actuator. To achieve practically meaningful actuation and sensing capabilities, a piezoelectric bimorph consisting of two piezoelectric layers is widely used T and ε = [x εy εz γyz γzx γxyε]T represent stress vector and strain vector, respectively. E = [Ex Ey Ez]T, the electric field vector, D = [Dx D y D z]T, the electric displacement vector, c, the elastic coefficient matrix, g, the dielectric coefficient matrix, and e, the piezoelectric stress coefficient matrix.A piezoelectric bimorph made of two identical PZT-4 piezoelectric layers, which has been investigated by Fernandes , is consu̅, υ̅, w̅ denote the displacements of an arbitrary point on the mid plane z = 0, α̅ and −β̅ denote the rotations of a transverse normal about the y and x axes, respectively. In the FSDT, the transverse shear strains are assumed to be constant with respect to the thickness coordinate. The constant state of transverse shear strains across the thickness is a gross approximation of the true strain field, which is at least quadratic through the thickness.An ESL model adopting the FSDT is adopted to describe the mechanical displacement. The displacement field of a piezoelectric bimorph based on FSDT takes on the form ,18:(2)uU is the displacement vector, and U̅ is a generalized displacement vector. Then We define:L is the derivation operator defined as:The infinitesimal strain components associated with the displacements are given by:2.2.e, is then mapped to a reference element, denoted by Ωref: ξ ∈ × η ∈ , using an invertible local mapping. The discretization procedure is illustrated in The Legendre polynomials based SE method can be described as follows: the bimorph is firstly discretized using a set of non-overlapping rectangular elements, as in the traditional FE method. Each rectangular element, denoted by Ωiξ, jη are defined in the local coordinate system ξ−η of the reference element Ωref as roots of the following polynomial expression:NP is the N-th order Legendre polynomial. In fact, the nodes are the 2-D Gauss-Lobatto-Legendre (GLL) points. In contrast to the classical FE method, the distribution of nodes is irregular, as shown in ref, as depicted in Subsequently, a set of nodes, denoted by iξ are defined as [ijδ denotes the Kronecker delta. The 2-D shape functions are constructed as a tensor product of the 1-D ones:The 1-D shape functions at the 1-D GLL points fined as :(10)hi is assumed to be linear across the thickness such that:Φ̃i is a column matrix composed of the electric potentials at the top and the bottom surfaces of the i–th sublayer, which can be expressed as:For the purpose of accurately modeling the distribution of the electric potential across thickness, each layer of the piezoelectric bimorph is subdivided mathematically into Φe is then introduced in the spectral plate finite element, which is defined as:In this way, the assumption of linear distribution of electric potential across the thickness is used not in the whole piezoelectric layer, but in each sublayer instead. As a result, the electric potential is approximated as piecewise linear across the thickness and it is expected that the quadratic distribution of the electric potential across the bimorph thickness can be approached with more sublayers adopted. As mentioned before, the mechanical displacement field is approximated using ESL model based on FSDT and the piezoelectric bimorph is discretized using 2-D mesh. To keep the compatibility, each sublayer of the piezoelectric layer is also discretized using the same mesh. Consequently, an element potential vector iϕ, is assumed to be constant over the element and ϕ0, ϕ1, ⋯ ϕ2n are treated as elemental DOFs, as illustrated in x, y. Thus the present method combines an ESL theory for the displacement and a piecewise linear approximation for the electric potential. Under the quasi-electrostatic approximation, the electric field and the electric potential in each sublayer have the following relationship:Ei(z) is the electric field of the i-th sublayer, The surface potential of the sublayer, 2.4.ρ is the mass density, Ps is the surface force vector, qs is the surface charge density vector. J is the well-known Jocobian matrix of the mapping (13) which is defined as:By applying Hamilton's principle, the elementary dynamic equations for the piezoelectric bimorph plate can be obtained:i.e., (1) sensor function with the top and bottom surfaces grounded and a uniform pressure load of S0 = 1,000N/m2 applied to the upper surface, and V0 = 50V applied to the top and bottom surface of the bimorph. By applying the electric boundary conditions, the DOFs for the electric potential are condensed out such that Muu, Kuu, Kp, Fu and Fa are the assembled counterparts of matrices q is the global nodal displacement vector. Since the DOFs for the sublayer electric potentials have been condensed out, this approach will not result in a large number of potential DOFs. For the purpose of static analysis, the governing equations in The GLL integration rule is then used to calculate the characteristic matrices and the nodal force vector in 3.i.e., the deflections of the bimorph and the distribution of the electric potential across the bimorph thickness. A simply supported rectangular piezoelectric bimorph shown in In this section, the derived SE model is converted into a numerical code and case studies are carried out to validate the effectiveness and the capability of the present model for predicting both the global responses and the local responses, a and width b of the bimorph are 25 mm and 12.5 mm respectively. Two values of slenderness ratio, S = a/h = 5 and, S = 50 which represent the thick and thin bimorph plate, respectively, are considered. Unless otherwise stated, the order of Legendre polynomial is chosen as 5, and the mesh in E0 is taken as E0 = 1010V/m. For the purpose of comparison, a coupled 3-D analysis is carried out using 20-noded hexahedral 3-D piezoelectric elements (C3D20RE) with a mesh size of 40×20×10 in ABAQUS and the results from the coupled 3-D FE analysis are taken as accurate.The length 3.1.S0 = 1,000N/m2 is applied to the upper surface and the bimorph is used as a sensor with the top and bottom surfaces grounded. The variations of both the deflection W and the electric potential Φ across the bimorph thickness at the centre of the bimorph plate  for the slenderness ratio S = 5 and S = 50 are shown in W estimated by the present method adopting different number of sublayers is constant through the thickness and it is a good approximation of the nonlinear distribution described by the coupled 3-D analysis. The present model based on FSDT with assumption of uniform deflection through the thickness cannot predict the nonlinear variation of W through the thickness. The electric potentials induced by the deformation of the bimorph through the direct piezoelectric effects are shown in W across the bimorph thickness, it may be able to provide good approximate results for Φ with appropriate number of sublayers for both thick and thin bimorph plates.For this case a uniform pressure load of 3.2.V0 = 50V is applied to the top and bottom surfaces of the bimorph with intermediate electrode grounded. The through-the-thickness variations of W and Φ at the centre of the plate for S = 5 and S = 50 are shown in To achieve practically meaningful actuation capabilities and guarantee that the piezoelectric material behaves linearly, an electric potential of W. Similar to the previous observation, the constant deflection W through the thickness calculated by the present method adopting different number of sublayers is a good approximation of the nonlinear distribution provided by the coupled 3-D analysis. It is noticed that as the sublayer number increases a smaller deflection is obtained which is also pointed out by Wang [S = 5 and S = 50, respectively. It can be observed that the almost linear distribution of Φ across the thickness predicted by the present method for both thick and thin bimorph plates is in excellent agreement with the coupled 3-D analysis, indicating that the nonlinear induced electric potential is insignificant compared to the externally applied potential. Consequently, the conventional linear electric potential assumption [Once again, the present model based on FSDT cannot predict accurately the through-the-thickness distribution of  by Wang . The elesumption  may be a4.The present work aims to develop an efficient SE model with electric potential DOFs for the static electromechanical response of a piezoelectric bimorph. The approach is the combination of an ESL model based on FSDT for the mechanical displacement with a sublayer model based on the piecewise linear approximation for the electric potential. 2-D GLL shape functions are used to discretize the displacements and then the governing equation of motion is derived following the standard SEM procedure. By applying the electric boundary conditions, the DOFs for the electric potential are condensed out such that the present model will not result in a large number of potential DOFs.i.e., a uniform pressure load applied to the top surface and a uniform potential applied to the top and bottom surfaces. To validate the effectiveness and the capability of the present model for investigation of both global and local response of the piezoelectric bimorph, the numerical results thus obtained are compared to those from 3-D analysis using ABAQUS. The results indicate that the deflection W estimated by the present method is a good approximation of the nonlinear distribution predicted by the coupled 3-D analysis. It is further shown that the present model provides very accurate prediction for the electric potential distributions across the bimorph thickness even for rather thick bimorph plate without introducing any higher-order electric potential assumptions. It is also revealed that the conventional linear electric potential model is accurate enough to predict the local electric potential response for the case of actuators. This observation consists with the previous findings proposed by Yang [W across the thickness is constant. Nevertheless, it is accurate enough to investigate the global response of the piezoelectric bimorph. The present work is important for researchers to better understand the nonlinear induced electric potential for bimorph sensor and actuator. An important extension of the present research is to study the vibration characteristics of the piezoelectric bimorph based on SE method. The influence of the induced stiffness matrix on the natural frequencies of the bimorph plate under various electric boundary conditions is to be investigated. The convergence study of the present model with respect to the order of the Legendre polynomial is also a practical and interesting problem to be conducted.Numerical simulations based on the present model are carried out for two different load cases,  by Yang . One of


Being born small for gestational age (SGA) is associated with decreased insulin sensitivity and increased blood pressure in childhood, but the association with clinical disease in early adulthood is less certain. The Danish Medical Birth Registry has registered all births in Denmark since 1973, but due to variable data quality, data is most often used only from 1981 onwards, and birth registers in other countries may have similar problems for the early years. We wanted to examine whether the data can be used for identification of children born SGA and used in future research.All persons born between 1974 and 1996 were identified in the Danish Medical Birth Registry (n = 1.704.890). Immigrants and children without data on gestational age and birth weight were excluded, and a total of 1.348.106 children were included in the analysis. The difference between the different variables used in the history of the registry were examined, and the quality of data in the birth registry from 1974-1981 was examined and compared to subsequent years.Data on birth weight and gestational age in the early years of the registry is inconsistent, and the identification of children born SGA is inaccurate, with 49% false-positives. The biggest source of error is due to the rough and inaccurate intervals used for gestational age. By using –3 standard deviations as a cut-off for the identification of children born SGA, the number of false-positives was reduced to 9%, while the amount of false-negatives were increased.Choosing –3 standard deviations for identifying children born SGA is a viable, though not optimal solution for identifying children born SGA. Overall the data in the registry is of sufficient quality to be used in further medical research. Children born small for gestational age (SGA) have previously been reported to have an increased risk of infant mortality, cerebral palsy, cardiovascular disease, insulin dependent diabetes mellitus (IDDM), non insulin dependent diabetes mellitus (NIDDM) and metabolic syndrome The Danish Medical Birth Registry offers a tool for epidemiological research on the population level. The Danish Medical Birth Registry has registered all births in Denmark since 1973, holding data on birthweight and gestational age among others. The data in the early years from 1973 to 1981, however, are imperfect, coding birthweight and gestational age into intervals, probably due to data storage being expensive at the time. Furthermore some birth cohorts have a substantial amount of missing data Several studies using The Danish Medical Birth Registry have excluded the early birth cohorts, some citing the inconsistencies as a reason The early birth cohorts are of interest first of all because they approach an age where the incidence of NIDDM and cardiovascular disease increases markedly and secondly because these birth cohorts are the largest in recent times in Denmark, including over 500.000 births from 1973 to 1981. Thus we examined whether they can be compared to subsequent birth cohorts and included in epidemiological research with a focus on smallness for gestational age. This is a very active research field and other early birth register cohorts may have similar limitations.Data were extracted in December 2006 from the Fertility Database held by The National Board of Health in Denmark. The registers include data on birthweight, gestational age, place of birth, and Apgar score. Data was extracted for a yet unpublished study on late effects of brain injury in neonates, and this is an opportunistic study using the same data.All children born in Denmark as well as legal immigrants are assigned a civil personal registration number (CPR number). The CPR number is a 10-digit number, with the 6 first digits being the date of birth, and the last four being a unique identifier for each person. The number is used for personal identification and registration in many public and private instances, and allows linking data from different registries in medical research.The data collection for the study was approved by the Danish Data Protection Agency. The CPR-numbers were encrypted and thus anonymised for the researchers. Under Danish legislation it is not necessary to apply for approval by The Danish National Committee on Biomedical Research Ethics for database studies as long as the study is approved by the Danish Data Protection Agency. Neither is it necessary to get written consent from individuals for database studies.We identified 1.704.890 people in the Danish CPR register born between 1974 and 1996. 1.545.641 persons were registered in The Danish Medical Birth Registry, the remainder primarily being immigrants (n = 159.249).We obtained the recorded birthweight and gestational age and individuals without data on both were excluded  as well as stillbirths . To exclude unlikely values, birthweight by gestational age (GA) was compared to the expected value as calculated by the use of Marsal's charts of normal intrauterine growth vegt1 in 250-gram intervals . In our analysis we assigned the centre value of the interval as the specific birthweight for all individuals.From 1974 to 1996 birthweight was coded in the variable vaegt was the primary birthweight variable. From 1979 to 1991 vaegt was coded in 10-gram intervals and afterwards in 1-gram intervals.From 1979 onwards, the variable fuldb, which corresponds to term/preterm birth. The value 0 is term birth , and the values 1–8 define increasing degrees of prematurity . The mean values and the difference between the two weight variables were calculated for each year and gestational week.A graphical representation of the 1-gram weight variable was made with complete histograms for each year, to assess whether there were preferences for certain values.When GA given in completed weeks was converted to GA in days, we multiplied by 7 and added 3 to approximate the mean number of gestational days for the week interval.th power model of birthweight per gestational age  for the years 1974 to 1977 to estimate a more correct distribution of gestational age for the different values.Expected birthweight was calculated using the sex-specific models published by Marsal The trend in weight per GA was graphed for the two GA variables. For each gestational week, development in birthweight was modelled with linear regression using the PROC GLM procedure.For the children born in 1979 we pooled the children born at GA 38–45 in one single group to examine the effect of the GA classification used from 1973 to 1977. We identified all children with a birthweight less than –2 and –3 SD of expected in both groups. Afterwards we compared the classification of children born as SGA in the pooled group with their classification using their true GA in completed weeks.In 1979 the birthweight was coded in 10-gram intervals, but there was a distinct preference for 100- and 50-gram values . In 1996The mean value of the 1-gram variable was higher than the 250-gram variable for all years, though the difference decreased from approximately 22 grams in 1979 to 16 grams in 1996.For intervals below the mean, the 1-gram value had a higher mean interval value than the 250-gram value, while the opposite was true for intervals above the mean (data not shown).From 1974 to 1977 the gestational age was coded in the birth register as seen in When correcting for gestational age in studies it is necessary to recode these values to a gestational age in days or weeks. When comparing the weight for the GA value  = 0 age to GA in weeks, it follows the trend in weight for week 40 . When doWe made a model of birthweight per GA in 1978 and calculated the corresponding week in the regression model when using the 1977 mean birthweight .In 1978 the procedure of reporting gestational age to the Danish Medical Birth Register changed. This resulted in a significant increase in the number of cases without data on GA. There are only small differences in the mean birthweight between cases with and those without data on GA . The disThe mean gestational age was reduced by approximately 2 days from 1978 to 1996, and using Pearson's test for trend for GA 34–39 we found a significant positive correlation between year and birth week frequency , whereas there was a strong negative correlation (p<0.0001) between year and frequency for GA 40.The percentage of children with a birthweight less than -2 SD of expected were calculated using the gestational ages for the nominal and the calculated model shown in The distribution in The total number of children classified as SGA, using the standard –2 SD as a cut-off, increased from 4.8% to 6.9% when the children were pooled. In the pooled group 49% of the SGA population were false positive SGA. When using –3 SD as a cut-off, the number of false positives was reduced from 49% to 9%. Those still false-positive had a mean SD of -1.8. The number of false-negatives was increased to 62% of the SGA population, and the total number of children classified as SGA was reduced from 6.9% to 2%. The mean SD was reduced with 0.25 SD from -2.47 SD to -2.72 SD.Our results show that data from the early years of the Danish Medical Birth Registry is of mediocre quality but can be used in further research with care.The problems are:A preference for 100- and 50-gram values when reporting birthweightA classification of gestational age in intervalsYears with many children missing data on gestational ageA small difference in mean birthweight in children with and without data on gestational ageDue to the first and second problem there are difficulties in classifying children as SGAThese finding are in accord with other studies using data from these years, though the approach to overcome them have differed The preference seen for 100- and 50-gram values in the early years could be due to the use of mechanical scales, and would thus be expected to disappear as digital scales were more frequently used. As the preference persisted in 1996 where most Danish hospitals had switched to digital scales, it seems to be due to rounding when reporting the weight, rather than limitations in the scales.The old 250-gram variable tended to underestimate the weight of children with a birthweight lower than the mean, as expected given the approximately normal distribution of birthweight. The underestimation was reduced from 22 grams in 1979 to 16 grams in 1996. Overall, using the 250-gram variable instead of the 1-gram variable increased the number of children regarded as SGA by about 15%. The difference is small, but must be taken into consideration if comparing data directly between the two birthweight variables.From 1978 to 1981 data on gestational age is missing on 13–30% of children. We found a small difference in birthweight between the groups, with a tendency for the children with missing data on gestational age being slightly heavier, though the distribution of birthweight or gestational age did not differ systematically between the two groups. Therefore it appears safe to consider the births with data on gestational age an unbiased sample of those birth years. The coding of gestational age in the years 1974 to 1977 was very insufficient. Not only was term birth coded in a very wide gestational age interval, coding practice was also inconsistent with the documentation of the Danish Birth Register. For preterm children the gestational age coding is more accurate.th percentile of the birth weight distribution as SGA. Comparing birthweight from 1974 to 1996 to Marsal's curves of birthweight, which are based on children born 1986–87, however, may not be optimal since the mean birthweight changes over time. From 1974 to 1986 the mean birthweight for GA 40 and 37 increased by 53 and 122 grams, respectively, corresponding to an underestimation of weight per gestational age of –0,12 and –0,25 SD if using the 1986 birth curves on the 1974 cohort. On the other hand, considering optimal growth as an absolute standard, at least some of this difference may be considered real.We chose to use a normal fetal growth standard with a -2 SD cut-off, since this is more likely to represent optimal/suboptimal fetal growth, instead of a cut-off classifying children below the 10The classification of SGA children was more severely obstructed by the coding of both birthweight and gestational age in wide intervals. This resulted in almost 50% of the population nominally SGA, being false positive. Using a threshold of –3 SD, however, identified 5000 individuals as being SGA of a cohort of over 270.000 Danes, aged 34–37 years at present. Only 9% of these 5000 individuals were false-positive SGA, with a mean SD of -1.8, and the rest had a birthweight that was truly less than – 2 SD for their gestational age. This SGA group, however, had a mean GA that is lower than expected and this bias may be a problem in contexts where gestational age is important.The mean SD in this group was 0.25 SD lower than expected, which could signify that the identified cohort is more severely growth restricted than expected using -2 SD as a cut-off.Danish national registries hold much medical, educational and social information, allowing large-scale epidemiological studies by linking the CPR numbers. A study using SGA as defined above in the birth cohorts from 1974 to 1977, on a condition with a diagnosed prevalence of 1%, would have a 91.9% probability of detecting a relative risk of 1.5  between this SGA group and the general population.The overrepresentation of individuals born between 37 and 40 weeks in the SGA-group will constitute a significant bias for the study of outcomes such a neonatal respiratory morbidity, other neonatal complications, as well as short- and long term neuropsychological outcomes, for all of which even modest prematurity is an important risk factor In conclusion, it is possible to identify a large cohort of young adults born SGA in the years 1974–1977. The cohort exhibits GA bias and has a mean SD that is -0.25 SD lower than expected. Given the expected bias, this cohort can not be used as a replacement for later and more precise birth cohorts, but may be a valuable supplement.


Background: Previous research suggests perfluorooctanoic acid (PFOA) and perfluorooctane sulfonate (PFOS) may be associated with adverse pregnancy outcomes.Objective: We conducted a population-based study of PFOA and PFOS and birth outcomes from 2005 through 2010 in a Mid-Ohio Valley community exposed to high levels of PFOA through drinking-water contamination.n = 1,630) to identify the outcomes of preterm birth (< 37 weeks gestation), pregnancy-induced hypertension, low birth weight , and birth weight (grams) among full-term infants.Methods: Women provided serum for PFOA and PFOS measurement in 2005–2006 and reported reproductive histories in subsequent follow-up interviews. Reported singleton live births among 1,330 women after 1 January 2005 were linked to birth records (n = 158) or low birth weight (n = 88). Serum PFOA and PFOS were both positively associated with pregnancy-induced hypertension (n = 106), with adjusted odds ratios (ORs) per log unit increase in PFOA and PFOS of 1.27  and 1.47 , respectively, but associations did not increase monotonically when categorized by quintiles. Results of subanalyses restricted to pregnancies conceived after blood collection were consistent with the main analyses. There was suggestion of a modest negative association between PFOS and birth weight in full-term infants , which became stronger when restricted to births conceived after the blood sample collection .Results: We observed little or no evidence of association between maternal serum PFOA or PFOS and preterm birth (Conclusion: Results provide some evidence of positive associations between measured serum perfluorinated compounds and pregnancy-induced hypertension and a negative association between PFOS and birth weight among full-term infants.http://dx.doi.org/10.1289/ehp.1206372Citation: Darrow LA, Stein CR, Steenland K. 2013. Serum perfluorooctanoic acid and perfluorooctane sulfonate concentrations in relation to birth outcomes in the Mid-Ohio Valley, 2005–2010. Environ Health Perspect 121:1207–1213;  Perfluorooctanoic acid  and perfluorooctane sulfonate (PFOS) are synthetic, environmentally persistent perfluorinated compounds (PFCs). PFOA has been used in the manufacture of fluoropolymers such as polytetrafluoroethylene  since the 1940s. PFOS exhibits similar properties and, like PFOA, has been used in a variety of consumer products  for its stain-, grease- and water-resistant properties. Biomonitoring of the general U.S. population indicates that exposure to PFOA and PFOS is nearly ubiquitous, with > 99% of people tested in 2003–2004 showing detectable levels of both PFOA and PFOS in their blood .Previous epidemiologic studies of PFCs and pregnancy outcomes in the general population provide inconsistent evidence of associations with birth outcomes such as birth weight and length of gestation ; both PFn = 32,254) participated in one or two follow-up interviews between 2008 and 2011 as part of the Community Follow-up Study  with residents in certain water distribution districts more highly exposed than others . The medup Study . For theup Study .Four recent retrospective cohort studies have examined relationships between PFOA and pregnancy outcomes in this highly exposed Mid-Ohio Valley region . Two stuStudy population. Participants who enrolled in the C8 Health Project between 2005 and 2006  completed a demographic and health questionnaire and provided serum for measurement of PFOA and PFOS. A subset of participants in the C8 Health Project  participated in one or two follow-up interviews between 2008 and 2011 as part of the Community Follow-up Study. To be included in this analysis, women must have provided a blood sample at enrollment in the C8 Health Project, completed at least one follow-up interview, and reported at least one live birth between 2005 and 2010. In the follow-up interviews (completed online or by trained telephone interviewers) women were asked about their entire reproductive history, including pregnancies that had occurred since their C8 Health Project enrollment interview. For each reported pregnancy, women were asked “Would you please tell us the month and year that the pregnancy ended?” and “What was the outcome of the pregnancy?” with the following response options: “live birth of single child,” “live birth of multiple children,” “miscarriage,” “stillbirth,” “termination/abortion,” “tubal,” “molar,” “ectopic.” In these follow-up interviews women were not asked specifically about birth weight, gestational age, or pregnancy-induced hypertension (PIH); therefore, these birth outcome data were obtained by linkage to birth records.n = 2012) were linked to birth records at the Ohio Department of Health  and the West Virginia Department of Health and Human Resources  using combinations of the following variables: mother’s social security number, mother’s first name, mother’s last name, mother’s date of birth, and month and year of the reported live birth. Of the live births reported in the follow-up interviews, 83%  were ultimately linked to a birth record in West Virginia or Ohio. Maternal serum concentrations were similar between the population of matched and unmatched births with average PFOA levels of 31.1 ng/mL (median = 12.9) in the unmatched group compared with 31.0 ng/mL (median = 14.3) in the matched group. Average PFOS levels were 15.3 ng/mL (median = 12.8) in the unmatched group and 15.6 ng/mL (median = 13.9) in the matched group. Other characteristics also were similar between the two groups (data not shown), except that women who were not matched tended to be more highly educated (38% were college educated vs. 23% in the linked group) . Birth record data for births occurring through 2004 were analyzed previously . Approxin = 27 births). Births to women > 45 years of age were excluded (n = 2). After exclusions, there were 1,630 singleton births included in the analyses.We examined the following outcomes, obtained from birth records, in relation to PFOA and PFOS: preterm birth (< 37 weeks gestation), PIH, low birth weight , and continuous birth weight (grams) among full-term infants (≥ 37 weeks gestation). Gestational age was calculated based on LMP. PIH is defined as the onset of hypertension after the 20th week of gestation; indication of PIH on the birth record may also include preeclampsia (PIH accompanied by protein in the urine). Analyses were restricted to singleton births among white women due to the small number of births to nonwhite women in the study  were measured at the time of enrollment. A woman was classified as diabetic if she reported on any survey that she had ever been told by a doctor she had diabetes, other than during pregnancy. Informed consent for participation in the enrollment health interview, blood collection, and release of medical records (separate forms) were obtained from study participants at enrollment by Brookmar Inc.  . ParticiExposure assignment. Blood sample processing and analytical methods for measurement of PFOA and PFOS in the serum have been described in detail previously  increase . Covariates included in all models were chosen a priori: maternal BMI at C8 Health Project enrollment based on standard categories   , years oOur primary models included natural log-transformed PFOA or PFOS concentrations as continuous exposures. We also modeled untransformed continuous concentrations and quintiles of PFOA and PFOS, and tested for trends across quintiles by including an ordinal variable for quintile in the model. The bottom (referent) quintile for PFOA would capture approximately 90% of U.S. adult women based on NHANES  in 2003–2004, consistent with elevated exposure in the study population due to the introduction of PFOA from the DuPont plant into the environment. In contrast, the bottom quintile for PFOS, which was from background sources only, would capture approximately 10% , perhapsn = 783, referred to as “first prospective births”) to avoid possible biases related to effects of ongoing or recent pregnancies on serum PFOA and PFOS concentrations. This analysis also excluded any future pregnancies among women who were pregnant at enrollment. We also conducted analyses excluding births to diabetic women, who have a higher baseline risk of many adverse pregnancy outcomes; analyses restricted to births that occurred in 2005–2007 only, which were closest in time to blood sample collection; analyses adjusting for both PFOA and PFOS in the same model; and analyses using alternative forms of model covariates, for example controlling for BMI using cubic terms instead of categories , and adjusting for birth year instead of time between conception and serum measurement. Additionally we conducted analyses of birth outcomes stratified by parity (first births to nulliparous women or births to parous women). All analyses were conducted using SAS version 9.2 software .For each outcome we conducted a variety of sensitivity analyses, including analyses restricted to each woman’s first  birth conceived after blood sample collection . The distribution of absolute concentrations (nanograms per milliliter) was similar between the two PFCs except for the top third of the distribution where PFOA concentrations were more right-skewed . Women with normal BMI, no previous births, or higher education at enrollment had higher PFOA and PFOS serum levels than other women.The distribution of maternal serum concentrations of PFOA and PFOS assigned to the 1,630 births and concentrations by maternal factors are shown in Preterm birth. The prevalence of preterm birth was 9.7% (n = 158). There was little evidence of an association between PFOA or PFOS serum levels and preterm birth in the overall cohort or when restricted to 783 births conceived after sample collection  (egnancy) . SensitiPregnancy-induced hypertension. There were 106 birth records (6.5%) indicating PIH; 30 records indicating chronic hypertension were excluded from this analysis. In the full cohort both PFOA and PFOS were associated with increased odds of PIH based on log-transformed and categorical analyses; we estimated a 27% increase in odds of PIH per log unit increase in PFOA  and a 47% increase in odds of PIH per log unit increase in PFOS  (p = 0.005), although ORs did not increase monotonically and were similar among quintiles 3, 4, and 5. Categorical analyses of PFOS also showed elevated odds of PIH relative to the lowest quintile without a monotonic trend (p-trend = 0.11). Analyses restricted to the first prospective pregnancy showed similar patterns of association but with less precision due to the limited numbers. The OR (per log unit) for PFOS and PIH was higher in these subanalyses .6, 104%) . Odds raResults of sensitivity analyses for log-transformed PFOA and PFOS and PIH  were generally consistent with the primary analysis for PFOA. For PFOS, the OR for PIH was reduced to 1.17  when analyses were restricted to births during 2005–2007, which were closest in time to the serum measurement, but more heavily weighted toward births that occurred before exposure assessment. Alternative approaches to covariate control had little impact on PIH ORs for PFOA or PFOS. When stratified by parity, ORs for a log unit increase in serum concentrations were virtually identical between births to nulliparous and parous mothers for PFOA , but not for PFOS . The OR for log-transformed PFOS was less precise but otherwise similar to the OR for all births when restricted to nulliparous births conceived after serum collection .Low birth weight. There were 88 births (5.5%) with a birth weight < 2,500 grams. There was little evidence of an association between serum PFOA or PFOS levels, although estimates were imprecise . Otherwise, sensitivity analyses were generally consistent with the primary analyses .analysis . There wp > 0.05) negative association between PFOS and birth weight among full-term infants. Subanalyses restricted to the births conceived after serum samples were collected were largely consistent with the main analyses, but with stronger positive associations between PFOS and PIH, and stronger, statistically significant negative associations between PFOS and birth weight. Although the analysis was limited in power, we were able to assess birth outcomes in relation to preconception biomarkers of exposure, in contrast with previous studies of birth outcomes and serum PFC concentrations measured in samples collected during or after pregnancy , but both compounds were associated with maternal BMI and parity. These associations may reflect effects of body weight and parity on the metabolism and excretion of PFCs, which raises concerns that associations between PFCs and PIH also may be a consequence of differences in pharmacokinetics between women at risk of developing PIH and other women. As others have noted, this kind of bias could occur even in the context of a prospective study  was within 3 years of the serum measurement, which is close to the average half-lives of PFOA (3.8 years) and PFOS (5.4 years) . AverageWhereas birth weight is thought to be accurately recorded on the birth record, gestational age is recorded with substantial error, and PIH is known to be underreported . In addiWe observed several expected associations with established risk factors that provided some degree of confidence in the outcome data. PIH showed expected associations with parity, obesity, and diabetes . PretermIn conclusion, we observed evidence of associations between serum PFOA and PFOS and PIH, and between PFOS and birth weight among full-term infants, but little evidence of associations with low birth weight or preterm birth in a community cohort highly exposed to PFOA. In light of previous findings in the Mid-Ohio Valley with regard to PFCs and hypertensive disorders of pregnancy, and the biological plausibility of such associations, further investigation with more refined outcome classification is warranted.(283 KBB) PDFClick here for additional data file.


The aim of this study was to validate a model for optimal birth weight derived from neonatal records, and to test the assumption that preterm births may be considered optimally grown if they are not exposed to common factors that perturb fetal growth.Weights of fetuses were estimated from serial biometric ultrasound scans  and combined with neonatal weights for a prospective pregnancy cohort (N = 691). Non-Caucasians, fetuses subsequently born preterm and those with diagnosed or suspected determinants of aberrant growth were excluded leaving fetuses assumed to have experienced normal growth. A generalised linear longitudinal growth model for optimal weight was derived, including terms for gestational duration, infant sex, maternal height and birth order. This model was compared to a published model derived solely from birth weights.Prior to 30 weeks gestation, the published model yielded systematically lower weights than the model derived from both fetal weight and neonatal weight. From 30 weeks gestation the two models were indistinguishable.The model for optimal birth weight was valid for births that have attained at least 30 weeks gestation. The model derived from both fetal and neonatal weights is recommended prior to this gestation. Standards for fetal weight are typically derived from birth weights of neonates born at different gestational ages ,2. HowevOptimal weight may be interpreted as a result of growth achieved in the absence of any factors that pathologically affect growth. We have previously reported a method of assessing the appropriateness of fetal growth using the proportion of optimal birth weight . With thThe aim of this study was to validate a model for optimal birth weight derived from neonatal records, and to query the assumption that preterm births may be considered optimally grown if they are not exposed to common factors that perturb fetal growth.Our analytic approach is summarised by the following steps:i. We selected a sample of optimally grown term births from a randomised controlled trial of serial biometric ultrasound scans in pregnancy. Weight was estimated from biometric measurements made during each ultrasound scan during pregnancy and obtained at the time of birth.US refers to optimal weights derived with this model.ii. A model for optimal weight was derived at each of these measurement occasions using the fetal and neonatal weights as the response variable and non-pathologic determinants of growth as explanatory variables. OWBW.iii. Optimal weight was then derived at each of these measurement occasions using the published model derived using only birth weights from an independent population, OWiv. The difference in optimal weight estimated by the two models was calculated at each week of gestation.A total of 9,222 serial ultrasound scans were obtained between 1989 and 1991 in Perth Western Australia for 2,860 pregnancies that resulted in a live birth  were appended to the ultrasounds , resulting in 3,539 measurements , femur length (FL), head circumference (HC) and biparietal diameter (BPD) measurements were obtained from the ultrasounds. Hadlock’s formula was applied to estimate fetal weight using AC, FL, HC and BPD measurements taken at or after 27 weeks gestation . In a coUS) derived from both the fetal ultrasound scans and neonatal measurements was determined using gestational age, maternal height, birth order and infant sex as predictors. All first order interactions between the predictor variables were considered along with polynomial forms of gestational age. Box-Cox transformations were applied to improve the fit of the model [A predictive model for the expected fetal weight  was calculated for each record using a previously published model that relied on birth weight measurements [The optimal weight , Third refers to the third birth (para 2), and More refers to the fourth birth or more. Height is maternal height measured during pregnancy.In that study, similar exclusion criteria were applied: multiple birth (3%), maternal smoking during pregnancy 27%), maternal vascular disease (7%), congenital anomaly (6%) and maternal pre-existing or gestational diabetes (4%) [%, maternBW from OWUS was calculated at each gestational age as:The proportional deviation of OWBW was considered an appropriate representation of OWUS at gestational ages where the 95% bootstrap confidence interval [The formula for OWinterval  for the Ethics approval was obtained by the Department of Health Western Australia, Human Research Ethics Committee.There was an approximately equal ratio of male to female births (51:49) Table . Mothersρ = 0.007191, SD = 0.001657).The Box-Cox procedure suggested the square root as the optimal transformation. The Pearson correlation between the observed and predicted fetal weights was 0.97. The square of this correlation, a pseudo R-squared, indicated that the model explained 95% of the variation in fetal weights. The autocorrelation in the residuals between consecutive days was low but statistically significant, and therefore systematic and non-ignorable (US (grams) expressed as:Parameter estimates are shown in Table where Time is the time of gestation represented in days since the last menstrual period.BW, OWUS, and proportional deviation of OWBW from OWUS were calculated from 16 to 39 weeks gestation because few ultrasound scans were conducted outside this interval  was non-zero (p < 0.0001) at 10 g (SD = 26 g) using fetometric ultrasound measurements taken before 27 weeks gestation. The mean difference between estimated fetal weight based on Scott’s model and the previously published model (Scott – OWBW) was also non-zero (p < 0.0001) at 197 g (SD = 68 g) using fetometric ultrasound measurements taken before 27 weeks gestation.Hadlock’s formula was applied to estimate fetal weight from ultrasound scans at or after 27 weeks gestation  and ScotOur study made use of a large prospectively collected sample of serial ultrasounds to derive an estimate of the optimal fetal weight and thereby also a method to ascertain the adequacy of fetal growth. We adopted a multi-faceted approach: weight was derived from ultrasound scans taken at multiple occasions during pregnancy and supplemented with weight measured at the time of birth, rather than sole reliance on birth weights and we excluded fetuses subsequently (i) born preterm, (ii) that died before 28 days of life, or (iii) that experienced pathologies affecting growth. We found that estimates of optimal weight based on a population of birth weights also subject to exclusions (ii) and (iii) but not (i) were systematically lower than fetal weight estimated using biometric ultrasounds prior to 30 weeks gestation.US and OWBW . Among this group, the recorded antepartum factors that might have contributed to growth restriction but were not excluded as common causes of growth restriction were: threatened abortion (antepartum haemorrhage before 30 weeks gestation), N = 13; urinary tract infection, N = 2; antepartum haemorrhage (not attributed to placenta previa or abruption), N = 39; asthma, N = 12; genital tract infection, N = 9; vaginitis, N = 4; significant psychological morbidity, N = 11; anaemia, N = 5; neoplasms, N = 7. Approximately 63% of this cohort experienced at least one of these factors. Among all term neonates in Western Australia, the prevalence of asthma was 9.7% and neoplasms  was 0.02%, which were both lower than the prevalence in this cohort. Further studies are required to confirm whether maternal asthma, cervical cancers and depression are disproportionately stronger risk factors for growth restriction at such early gestations in other populations. It remains to be demonstrated that preterm births, particularly very preterm births, can be considered optimally grown if they are not exposed to common factors that perturb fetal growth.It is plausible that the difference between OWW Figure  suggestslow-risk preterm births before 28 weeks gestation. This meant that a formal validation of the model used to estimate fetal weight from ultrasound measurements could not be conducted. However, we confirmed that results were not sensitive to the choice of Hadlock’s versus Scott’s method to estimate fetal weight from ultrasound measurements. Before 27 weeks gestation, Scott’s method and Hadlock’s method to estimate fetal weight from ultrasound measurements differed by only 10 g, whereas the difference between estimates obtained from Scott’s method and the model for optimal weight derived from birth weight measurements was almost 200 g.Although it is plausible that there is an increasing proportion of unusual causes of growth restriction with decreasing gestation of delivery before 28 to 30 weeks gestation, there are other explanations for the findings. Radiographers may have systematically over or under-estimated fetal biometric measurements. However, the measurements were taken by a limited number of experienced radiographers at a tertiary obstetric hospital. While both Hadlock’s and Scott’s models were selected because they performed favourably compared to alternative formulae, it is possible that both of these methods used to estimate fetal weight from ultrasound measurements systematically over-estimated true fetal weight in our study. Few ultrasound scans were taken within a week of birth among early BW was developed predominantly on a sample of births from 33 weeks gestation whereas the model for OWUS was based on measurements from ultrasounds that start typically from much earlier gestations  who reported that the median of the fetal weight distribution provided an upper bound for the median of the birth weight distribution between 25 and 35 weeks of gestation . HoweverThe methodology that we applied differs from those suggested by others in that centiles and z-scores were not produced . HoweverIt remains to be demonstrated that preterm births can be considered optimally grown if they are not exposed to common factors that perturb fetal growth. The model derived from both fetal and neonatal weights in this study is recommended before 30 weeks of gestation. From 30 weeks gestation the two models were indistinguishable.The authors declare that they have no competing interests.GP participated in the design of the study, performed the statistical analysis, and wrote the initial manuscript and later revisions. EB conceived the study, participated in its design and helped to draft and revise the manuscript. DL participated in the design of the study and helped to revise the manuscript. All authors read and approved the final manuscript.


Facial cleft deformities, including cleft lip with or without cleft palate (CL/P) and cleft palate (CP), are common congenital birth anomalies, especially in Asia. This study aimed to analyze the prevalence of CL/P and CP and to identify associated factors in Taiwan.This population-based epidemiological study retrospectively analyzed birth data obtained from the Department of Health in Taiwan for years 2002–2009. Frequency distribution, percentages and related predictors were investigated, and findings were presented by types of cleft deformities. Logistic regression analysis was performed to identify factors associated with cleft deformities.Overall prevalence of cleft deformities among 1,705,192 births was 0.1% for CL/P and 0.04% for CP over the 8-year study period. Higher prevalence of CL/P or CP was observed with multiple pregnancies, being male for CL/P, being female for CP, gestational age ≤37 weeks and lower birth weight (<1.5 kg). Both CL/P and CP were significantly associated with gestational age <37 weeks and birth weight<1.5 kg . CL/P was significantly associated with multiple parities (P = 0.0004–0.002). Male newborns and female newborns were significantly associated with CL/P and CP, respectively (both P<0.0001).Overall prevalence for congenital cleft deformities in study subjects was 0.1%, in keeping with high rates in Asia. Results suggest the need for awareness and early identification of those at high risk for cleft deformities, including newborns with gestational age <37 weeks, weighing <1.5 kg at birth and women with multiple parities, as a potential strategy to counter long-term adverse effects on speech and language in this population. Facial cleft deformities, including cleft lip with or without cleft palate (CL/P) and cleft palate (CP), are among the most common congenital birth anomalies. While the worldwide prevalence of such deformities is about 1.5 per 1,000 live births, the rate varies six-fold for CL/P and three-fold for CP Even when newborns with cleft deformities receive appropriate treatment, some still have facial deformity and speech impairment We conducted a population-based epidemiologic study utilizing retrospective secondary database analysis of national data collected by the Taiwan Bureau of Health Promotion of the Department of Health, Executive Yuan from 2002 to 2009. The Executive Yuan collects health information from all hospitals, as well as other providers, to help guide the administration of public health policy in Taiwan. For this analysis, we used data from the Birth Registration Database Statistical analyses was done to examine the correlation of demographic characteristics of mothers and newborns with CL/P and CL. To examine the effect over time, the entire duration was divided into earlier and later periods (2002–2005 and 2006–2009), and the distribution of CL/P and CP was compared between the two groups using Chi-square test; data analyzed included maternal age , parity , newborn gender, gestational age  and birth weight . Count data were described using frequency and percentage. Logistic regression analysis was applied to identify factors associated with CL/P and CP. Statistical analysis was performed with SPSS version 18.0 software . A p value of <0.05 was established as statistical significance.The demographic and clinical characteristics of mothers and newborns are presented in We then categorized the information into two periods, 2002–2005 and 2006–2009, to determine the demographic characteristics related to prevalence of CL/P and CP, including mothers’ age, parity, gender, gestational age, and birth weight. Results from analysis are shown in Logistic regression analysis of the correlation between CL/P, CP and cohort demographic characteristics  showed tOur study characterizes the current national birth prevalence of CL/P and CP in Taiwan. We found overall birth prevalence of 0.1% for CL/P and 0.04% for CP over this 8-year period. Previous studies showed a worldwide prevalence of 0.1% for CL/P In our study, CL/P was more frequently found in and associated with male newborns. In addition to higher prevalence of cleft deformities, boys suffer greater psychosocial disruption from facial cleft deformities than girls In the present study, mothers with age ≥34 years were associated with an increased risk of having infants with a facial cleft deformity, but the association did not reach statistical significance. Other studies report similar results involving older mothers or parents We found that gestational age <37 weeks was associated with greater birth prevalence of facial cleft deformity, especially in the earlier cohort. Association of facial cleft disorders with low birth weight, premature birth or other developmental anomalies has led to calls for future research in prevention intervention and identifying genetic risk and predisposition In Taiwan, low birth weight of <1.5 kg was also significantly associated with facial cleft deformities. While birth weight is often not recorded in these studies, many find associated pregnancy complications  or other genetic anomalies in these infants, perhaps diagnosed prenatally Our study has certain limitations. The database did not provide information on maternal health or peri-conceptual conditions, which may play a role in development of facial cleft deformities In conclusion, overall prevalence for congenital cleft deformities in Taiwan was 0.1%, in keeping with high rates in Asia. Results of this study suggest the need for increased physician awareness, early identification of and possibleearly intervention for high risk mothers aged ≥34 years, those with multiple parities and those whose newborn weighed <1.5 kg at birth, which may be a potential strategy to counter long-term adverse effects on speech and language in this population.


Background: Although hyperthermia is a recognized animal teratogen and maternal fever has been associated with birth defects in humans, data on the relationship between high environmental temperatures and birth defects are limited.Objective: To determine whether pregnancies are potentially vulnerable to the weather extremes anticipated with climate change, we evaluated the relationship between extreme summer temperature and the occurrence of birth defects.Methods: We performed a population-based case–control study by linking the New York State Congenital Malformations Registry to birth certificates for the years 1992–2006. We selected nonmalformed infants from a 10% random sample of live births as controls. We assigned meteorologic data based on maternal residence at birth, summarized universal apparent temperature  across the critical period of embryogenesis, and estimated adjusted odds ratios (aOR) and 95% confidence intervals (CI) with multivariable logistic regression, controlling for confounders available on the birth certificate.Results: Among 6,422 cases and 59,328 controls that shared at least 1 week of the critical period in summer, a 5-degree increase in mean daily minimum UAT was significantly associated with congenital cataracts . Congenital cataracts were significantly associated with all ambient temperature indicators as well: heat wave, number of heat waves, and number of days above the 90th percentile. Inconsistent associations with a subset of temperature indicators were observed for renal agenesis/hypoplasia (positive) and anophthalmia/microphthalmia and gastroschisis (negative).Conclusions: We found positive and consistent associations between multiple heat indicators during the relevant developmental window and congenital cataracts which should be confirmed with other data sources. Hyperthermia is a well-known animal teratogen, and maternal fever has been associated with birth defects in human studies . In all There is a paucity of data on the teratogenicity of external heat-generating sources or hot environments, which may increase maternal core temperature during pregnancy, and the exposure assessments for these studies are typically based on postpartum maternal interviews about environmental conditions during early pregnancy. Existing data support a possible association of neural-tube defects with hot tub and sauna use during early pregnancy , but notIt is likely that temperatures will increase and extreme weather events will become more common with climate change . BecauseStudy population and data sources. The source population was all live births to residents of upstate New York  from 1992 through 2006. Appropriate institutional review board approvals were granted to access New York State birth certificate data from Vital Records  and birth defects data from the CMR. The birth certificate contains data on maternal and infant characteristics, such as maternal age, race/ethnicity, education, date of last menses, prenatal care, and maternal behavioral characteristics , infant date of birth, sex, birth weight, and gestational age. Study design and data linkage. We used a case–control study design to examine the association between high ambient temperature during the critical period of embryogenesis and the prevalence of birth defects. Cases were upstate New York live births with birth defects for the 1992–2006 study period, and controls were a 10% random sample of upstate New York live births without birth defects for the same period. Individual birth defects records were linked with birth certificate records.Outcome assessment. Using International Classification of Diseases, 9th Revision, Clinical Modification , education level , race , ethnicity , total previous live births , prenatal care , smoking , alcohol consumption , multiple birth , and infant sex  were derived from the birth certificates and considered as potential confounders and effect modifiers. Additionally, weather region was categorized based on maternal residential address.Statistical analysis. For the subset of births with at least 1 week of the critical period of embryogenesis within summer, we used unconditional logistic regression to estimate the association between each heat exposure indicator and selected birth defects group. Crude prevalence ORs and 95% confidence intervals (CIs) were calculated, and confounding and interaction were assessed using stratified analyses. Multivariable unconditional logistic regression was used to calculate adjusted ORs (aORs) and 95% CIs. Potential confounders were dropped out of the model in a reverse fashion, and a change in the effect estimate of > 10% was used as the criterion for keeping the variable in the model. Final models were adjusted for year, weather region, maternal age, ethnicity, education level, adequacy of prenatal care, and smoking status. For the associations between heat waves and each birth defects group, we performed stratified analyses by weather region, maternal race, ethnicity, age, education, adequacy of prenatal care, and smoking status. In our analyses, we used a p-value of 0.05.To control for potential detection bias due to increased surveillance among preterm infants, we performed subanalyses among term infants for congenital cataracts  and certA total of 6,422 selected birth defects cases and 59,328 controls that shared at least 1 week of the critical period of embryogenesis in summer were included in the analyses of upstate New York resident live births in the 1992–2006 study period. Distributions of maternal age, education, race, total previous live births, adequacy of prenatal care, smoking, alcohol use, multiple birth, low birth weight, preterm birth, region, and year of birth were significantly different between cases and controls . DistribA 5-degree increase in the mean daily minimum UAT in the critical period was significantly associated with an increased occurrence of congenital cataracts  and renal agenesis/hypoplasia  and a reduced occurrence of anophthalmia/microphthalmia  . AssociaCongenital cataracts were significantly associated with heat waves , number of heat waves , and number of days above the 90th percentile  . The preEstimates and 95% CIs from stratified analyses of heat waves by weather region, maternal race, ethnicity, age, education, prenatal care adequacy, and smoking status (data not shown) were imprecise and overlapping due to small sample sizes within strata. The only exception was associations between heat waves and cleft lip with or without cleft palate that varied significantly by ethnicity: Hispanic  versus non-Hispanic .We performed a subanalysis among term infants to assess the potential influence of detection bias in preterm infants (data not shown). Adjusted ORs for congenital cataracts remained statistically significant: 5-degree increase in mean daily minimum UAT , 5-degree increase in mean daily mean UAT , 5-degree increase in mean daily maximum UAT , heat wave , number of heat waves , and number of days above the 90th percentile . Associations with mild heart defects were similar to the full analyses, with no statistically significant associations.We further evaluated associations with congenital cataracts according to gestational week in the first trimester . AdjusteIn our population-based study of 15 years of birth defects data in upstate New York, we found elevated ORs for associations between multiple indicators of increased summer temperature and extreme heat during the critical period of embryogenesis and congenital cataracts. We also observed inconsistent positive associations between some heat indicators and renal agenesis/dysplasia, and inconsistent negative associations for some heat indicators with anophthalmia and gastroschisis.We did not see associations with other birth defects that have been associated with maternal hyperthermia in the animal and human literature. In our literature review of birth defects related to hot environments in early pregnancy, we found one study of neural-tube defects and two of congenital heart defects. Cooking in a hot kitchen; working, walking, or running in the sun; or lack of air conditioner use at home or work was not associated with neural-tube defects in a case–control study of 175 Mexican-American mothers of infants with neural-tube defects and 221 mothers with normal live births in a Texas–Mexico border population . CongeniCongenital cataracts. Congenital cataracts are opacifications of the lens of the eye that interfere with normal development of vision in infants and are a leading cause of preventable blindness and visual impairment in children  is considered normal, and 39°C (102°F) is regarded as the threshold for potential damage. Heat is teratogenic only when the temperature elevation coincides with a susceptible stage of development; and the type of defect depends on the particular stage of development , which will allow future comparisons and collaborations with other states participating in the National Environmental Public Health Tracking Network , and thermoregulating behaviors  were not taken into account. Several studies have shown that air conditioner use modifies estimated effects of heat-related mortality  would yield approximate adjusted CIs for congenital cataracts that include the null value . However, the associations with congenital cataracts are biologically plausible, particularly given stronger associations during the relevant developmental window of lens development, and associations were consistent across exposure metrics, making chance a less likely explanation for these findings.Last, because we performed multiple tests to examine the relationships between 28 birth defects groups and various heat exposure indicators in this hypothesis-generating study, statistically significant findings may have been attributable to chance. Under the null hypothesis, we would expect 4 of the 84 effect estimates displayed in Conclusions. We found positive and consistent associations with congenital cataracts of multiple ambient heat exposure indicators, including 5-degree increases in the mean daily UAT , a heat wave episode, the number of heat waves, and the number of days above the 90th percentile of UAT. Furthermore, these associations were strongest during the most relevant developmental window for congenital cataracts. Given our findings, associations with maternal fever reported by