Background. The neonatal immune system is not fully developed at birth; newborns have adequate lymphocytes counts but these cells lack function. Objective. To assess the activity of T-cells and the influence of the main perinatal factors in very preterm infants (birth weight < 1500 g). Design. Blood samples from 59 preterm infants (21/59 were dizygotic twins) were collected at birth and at 30 days of life to measure CD4+ T-cell activity using the ImmuKnow™ assay. Fifteen healthy adults were included as a control group. Results. CD4+ T-cell activity was lower in VLBW infants compared with adults (p < 0.001). Twins showed lower immune activity compared to singletons (p = 0.005). Infants born vaginally showed higher CD4+ T-cell activity compared to those born by C-section (p = 0.031); infants born after prolonged Premature Rupture of Membranes (pPROM) showed higher CD4+ T-cell activity at birth (p = 0.002) compared to infants born without pPROM. Low CD4+ T-cell activity at birth is associated with necrotizing enterocolitis (NEC) in the first week of life (p = 0.049). Conclusions. Preterm infants show a lack in CD4+ T-cell activity at birth. Perinatal factors such as intrauterine inflammation, mode of delivery, and zygosity can influence the adaptive immune activation capacity at birth and can contribute to exposing these infants to serious complications such as NEC. Although the immune system development begins early during fetal life, its maturation is not completed at birth as confirmed by the increased susceptibility of newborns and preterm infants to infectious diseases. The immune system of the fetus/newborn protects the infant against infection at the maternal-fetal interface but must also avoid the potentially harmful proinflammatory/Th1-cell response that can induce a dangerous reaction between mother and fetus. Thus, the suppression of proinflammatory response helps the infant in the transition from the intrauterine environment to the foreign antigen-rich environment of the outside world. Therefore this ineffectiveness that has been interpreted for a long time as a deficiency of the immune system could represent a biologically advanced response. in utero. This is confirmed by the decreased number of memory T- and B-cells and the increased number of naïve T- and B-cells in the neonatal bloodstream [The faults of the neonatal immune system are mostly related to a functional deficit of their components. Newborns and especially preterm infants have higher leukocyte and lymphocyte counts compared to adults ; howeverodstream . in vitro nonspecific stimulation with phytohemoagglutinin (PHA) as marker of CD4+ T-cell activity; therefore it is one of the methods for non-pathogen-specific immune monitoring. Exposing T-cells to a mitogenic stimulus such as PHA leads to their metabolic activation and polyclonal expansion, a process in which the ATP synthesis and release precede surface receptor expression, cytokine production, and other subsequent events. Therefore, the iATP levels offer a proxy for the degree of functionality of the cell-mediated immune response [In order to measure the immune cell function during the third trimester of gestation, the immune status in a group of preterm infants was evaluated at birth and after 30 days of life using the ImmuKnow assay . This immunological test measures the levels of intracellular adenosine triphosphate (iATP) afterDespite that, there are limited reports on the use of ImmuKnow assay in a pediatric population. Hooper et al. tested ImmuKnow assay in healthy children aged 1–17 years. The authors found that children aged 12 years and older have immune function levels equal to adults, while children aged less than 12 years present lower immune function levels compared to adults and older children .However, to the best of our knowledge, ImmuKnow assay has never been tested in newborns and preterm infants. in vitro stimulation with PHA in order to assess the basal condition of the adaptive immune system at birth, its development in the first month of life, and the influence of the main perinatal factors on the immune response in a population of very preterm infants.The aim of this study was to estimate the peripheral blood CD4+ T-cell activation rate in response toA prospective longitudinal study was carried out between November 2013 and July 2015 at the Neonatal Intensive Care Unit (NICU) of St. Orsola-Malpighi University Hospital in Bologna, Italy.All the infants with gestational age (GA) ≤ 30 weeks and birth weight (BW) < 1500 g admitted at birth to the NICU were considered eligible to the study. Infants with congenital malformations or congenital infections or born to a mother with pregnancy complications  were excluded. Before enrollment in the study, written informed consent was obtained from each of the infants parents.Fifteen healthy adults were also included as a control group. in vitro. Anamnestic and clinical data were prospectively collected during the hospital stay.Peripheral blood samples were collected in the first day of life and at 30 days of life from each patient to evaluate the pattern of lymphocytes subpopulations and the levels of iATP in activated CD4+ T-cellsSepsis was defined as presence of clinical signs of infection , elevation of infections markers , and ≥1 positive blood cultures.Necrotizing enterocolitis (NEC) was defined according to Bells modified criteria .The study was approved by the St. Orsola-Malpighi University Hospital Research Ethics Committee .CD4+ T-cell immune response was measured using the ImmuKnow assay . The test was performed at the Operative Unit of Clinical Microbiology—Laboratory of Virology—of St. Orsola-Malpighi University Hospital in Bologna, Italy, according to the package insert.μL of whole blood was diluted with sample diluent, added to wells of a 96-well microtiter plate , and incubated for 15–18 h with PHA in 37°C in 5% CO2. Magnetic particles coated with anti-human CD4 antibodies were introduced to the wells, and using a strong magnet, CD4+ T-cells were positively selected and separated. Then, the cells were lysed to release intracellular ATP. The released iATP was measured using luciferin/luciferase reagent and a luminometer.Peripheral blood was collected in sodium heparin tubes and processed on the same day. Briefly, 250 The patient's CD4+ T-cell immune response was expressed as the amount of iATP (ng/mL).In order to check each assay run, a control specimen collected from a healthy adult individual  was processed with patient specimen. Quality control samples were not included in the adult control group.The lymphocyte subsets were evaluated using the BD FACSCalibur Flow Cytometr  according to the manufacturer's instructions.p value < 0.05.Statistical analysis was performed by IBM SPSS . Data distribution was checked for normality with the Shapiro-Wilk test. As the data is not normally distributed, nonparametric tests were used. Univariate analysis was performed in order to evaluate which clinical variables were related to iATP values at birth and at one month of life: Mann–Whitney and Kruskall-Wallis tests were used for categorical variables and Spearman correlation test for continuous variables. Regression multivariate analysis was performed using as independent variables all those variables that proved to be significant in the univariate analysis. Statistical significance was defined as a Seventy-three eligible infants were admitted to NICU during the study period. Fourteen infants  were excluded because they either fulfilled the exclusion criteria  or peripheral blood samples could not be collected within the first day of life (4 infants died and 2 were transferred from our unit in the first days of life). The study was carried out in a population of 59 preterm infants with GA ≤ 30 weeks and BW < 1500 g; 21/59 were dizygotic twins (the study population does not include monozygotic twins).Samples were obtained at 30 days of life from 39/59 (66.1%) of the recruited newborns; the remaining 20/59 (33.8%) encountered death (4 infants) or were transferred to other hospitals before 30 days of life.p = 0.831) a significant positive correlation was observed at 30 days of life (p = 0.011).The pattern of lymphocytes subpopulations at birth and at 30 days of life is reported in p = .142), both these values were significantly lower compared to adults controls (p < 0.001) as shown in While there were no significant differences in CD4+ T-cells ATP levels at birth versus 30 days of life , and use of prenatal steroids .p = 0.005); this difference was no longer significant at 30 days of life (The twenty-one twins showed significantly lower median ImmuKnow iATP levels at birth compared to the remaining 38 singleton infants ( of life .p = 0.031); no difference was found at 30 days of life ; this difference disappeared at 30 days of life (Fifteen out of 59 preterm infants were born to mothers with pPROM (prolonged Premature Rupture of Membranes > 18 hrs); these infants showed a higher activity of CD4+ T-cells at birth compared to the 44 remaining infants  and twin pregnancies  were found to be independently associated with ImmuKnow iATP levels at birth .A multivariate analysis was performed including those variables (pPROM and twin pregnancy) which proved to be significant in the univariate analysis: both pPROM  (p = 0.036). On the contrary, the difference of median iATP levels between infants who did and did not develop NEC in the 25–27 weeks group was not significant, but we can not exclude the possibility that this is a consequence of the little data we had available.Five out of 59 (8.5%) preterm infants developed NEC ≥ stage 3 during the first week of life; these infants had significantly lower median iATP values at birth compared to the 55 infants without NEC (= 0.049) . We alsoNo difference in CD4+ T-cell activity was found at 30 days of life in relation to NEC condition.We found no influence of CD4+ T-cell activity on short-term outcomes , at birth or at 30 days of life.The neonatal immune system is not fully developed at birth and newborns are therefore exposed to the risk of infection by a large number of viruses, bacteria, protozoa, and fungi. This weakness can be partly attributed to the lack of preexisting immunological memory and competent adaptive immunity. Newborn infants have deficiencies in T-cell activation and cytokine production, B-cell immunoglobulin production, and interactions between T- and B-cells, when compared to adults . T-lymphFrom 19 weeks of gestation, T-cell subpopulations gradually increase in number and continue to rise after birth to peak at about 6–9 months of life. The numbers subsequently decline, to reach adult levels at 6-7 years of age . In term in vitro stimulation with a nonspecific mitogen such as PHA [μL).In this study, we used the ImmuKnow assay in order to investigate the function of CD4+ T-cells at birth in a population of preterm infants. This test is one of the methods for non-pathogen-specific immune monitoring and measures the levels of ATP produced by CD4+ T-cells in response toh as PHA . At thish as PHA –15 suggeOur findings show similar leukocyte and CD4+ T-cells counts to previously reported studies in preterm population , 16.We documented a lack of association between the number of CD4+ T-cells at birth and their function (measured by intracellular ATP production) and a positive correlation between number and activity of CD4+ T-cells at 30 days of life. Whether there is or there is no correlation between the number of lymphocytes and their function is still a matter of debate in literature.r = 0.24), emphasizing the usefulness of ImmuKnow assay to provide a quantification of CD4+ T-cells activity that is independent of cells absolute number.Kowalski et al.  demonstrHowever, other studies performed afterwards questioned this observation since a positive association between white blood cell count and ImmuKnow levels was observed , 18. DhaBecause each study is heterogeneous from the other, differing in characteristics such as population, type of organ transplant, immunosuppression protocols, and timing of ImmuKnow assay it is difficult to come to conclusions.We hypothesized that the correlation in our study is initially absent as a consequence of the immunological immaturity at birth  and becomes significant over time with the functional maturation of lymphocytes. Other neonatal confounding factors may also play a role and cannot be excluded and larger studies on immunocompetent subjects may validate our hypothesis.In our study the adult control population includes healthy HIV-seronegative immunocompetent people, whose lymphocytes subpopulations related data are not available.α, IFN-α, IL-4, IL-5, IL-10, IL-15, and IFN-γ in preterm infants' blood compared to adults [We found that preterm infants have a reduced CD4+ T-cell activation capacity compared to adults: the iATP values at birth are low, and despite a trend towards higher values over time, they remain significantly lower at 30 days of life. It is known that newborns, especially preterm infants, have deficiencies in both innate and adaptive immunity and many studies have demonstrated lower concentrations of cytokines such as TNF-o adults –23. HoweIn this study twin infants showed median ImmuKnow iATP levels at birth much lower compared to singletons. This is an interesting find and literature lacks information on the immunological peculiarities of multiple pregnancies. All the twins included in the study were dizygotic and thus immunologically different. We hypothesize that, in addition to a generalized status of immunological deficiency or insufficiency, the copresence of twins in utero may induce a deeper immune tolerance that involves both fetuses and the mother in order to avoid the potentially harmful immune reaction between the three. Our hypothesis is in accordance with previous studies that documented higher levels of Th2-cytokines in the blood of mothers carrying twins compared with singleton pregnancies  underlinγ, indicating a potential shift from Th2 to Th1 of the fetus [We found that pPROM significantly increases CD4+ T-cells activation at birth. This data is consistent with previous findings that reported that lymphocytes are activated during infections in utero, indicating that fetal adaptive immune response is at least partly responsive . It has he fetus , 30.We have also demonstrated that CD4+ T-cell activity at birth is increased in infants born after a vaginal delivery. Increasing evidence suggests that parturition itself is an inflammatory event .Our findings support the current knowledge that intrauterine inflammation/infection can lead to immune maturation in the fetus. But these differences disappear at 30 days of life most likely because all the categories  are exposed to the same extrauterine environment.In our study population, the levels of iATP at birth did not correlate with the risk of sepsis. This is not the first study that fails to detect an association between low ImmuKnow iATP values and infection. Regarding this topic, previous studies performed in immunocompromised adults showed conflicting results . In factSignificant positive correlations were observed between low CD4+ T-cell activation capacity at birth and NEC development in the first week of life. Although the etiopathogenesis of NEC is still a matter of debate, some authors support the involvement of innate immune system . We hypoThis study has some limitations. The study sample is small and lacks control groups both at term and with gestational age > 30 weeks.Another limitation is the absence of an immunophenotypic classification of CD4+ T-cells. The evaluation of Treg cells is crucial to understanding the immunological characteristics of the preterm infants. These CD4+ T-cells are provided with immunosuppressive functions and represent a high proportion of lymphocytes at birth with a significant inverse correlation with GA . The rolIn conclusion, preterm infants show a lack in CD4+ T-cells activation and fail to show a functional maturation of lymphocytes over the first month of life. An impaired ability to respond to stimulation can contribute to expose these infants to serious complications such as NEC. However, the adaptive immune response can at least partially be elicited during the fetal life by events occurring before delivery such as pPROM or vaginal delivery. Further studies in larger populations are needed to clarify these results and to better understand the cellular mechanism that regulates neonatal adaptive immune response to pathogens.


The Cessation in Pregnancy Incentives Trial (CPIT), which offered financial incentives for smoking cessation during pregnancy showed a clinically and statistically significant improvement in cessation. However, infant birth weight was not seen to be affected. This study re-examines birth weight using an intuitive and a complier average causal effects (CACE) method to uncover important information missed by intention-to-treat analysis.p < 0.0001). For this analysis, randomised groups were split into three theoretical sub-groups: independent quitters - quit without incentives, hardened smokers - could not quit even with incentives and potential quitters - required the addition of financial incentives to quit. Viewed in this way, the overall birth weight gain with incentives is attributable only to potential quitters. We compared an intuitive approach to a CACE analysis.CPIT offered financial incentives up to £400 to pregnant smokers to quit. With incentives, 68 women (23.1%) were confirmed non-smokers at primary outcome, compared to 25 (8.7%) without incentives, a difference of 14.3%  was 3338 g compared with potential quitters in the control group (who did not quit) 3193 g. The difference attributable to incentives, was 3338 – 3193 = 145 g . The mean difference in birth weight between the intervention and control groups was 21 g, and the difference in the proportion who managed to quit was 14.3%. Since the intervention consisted of the offer of incentives to quit smoking, the intervention was received by all women in the intervention group. However, “compliance” was successfully quitting with incentives, and the CACE analysis yielded an identical result, causal birth weight increase 21 g ÷ 0.143 = 145 g.Policy makers have great difficulty giving pregnant women money to stop smoking. This study indicates that a small clinically insignificant improvement in average birth weight is likely to hide an important clinically significant increase in infants born to pregnant smokers who want to stop but cannot achieve smoking cessation without the addition of financial voucher incentives.ISRCTN87508788. Registered on 1 September 2011.ISRCTN Registry, The online version of this article (doi:10.1186/s13063-017-2053-x) contains supplementary material, which is available to authorized users. Healthy birth weight is important for both early and long-term infant and child health as well as adult health. Healthcare costs are increased throughout life for low birth-weight babies. Babies born smaller develop more problems in the first month of life, requiring care in level 3 and level 4 neonatal intensive care units . Small bHealthcare costs associated with lower birth weight are substantial and are at least partially avoidable if women do not smoke during pregnancy . SmokingInterventions to help women to reduce or stop smoking during pregnancy, whilst of varying efficacy, have led to an overall moderate but statistically significant increase of 41 g in mean birth weight . Financip = 0.67).The Cessation in Pregnancy Incentives Trial (CPIT)  Additio and 2 waThis paper looks again at birth weight data from the CPIT and puts forward an alternative analysis. It aims to show the real effect on birth weight in women who would not quit without financial incentives but who manage to quit when financial incentives are added to routine cessation services. This intuitive approach is compared with a Complier Average Causal Effects (CACE) analysis.The CPIT was a phase II single-centre efficacy trial that offered up to £400 of financial incentives to self-reported pregnant smokers to engage with smoking cessation services and quit smoking near the end of pregnancy. Enrolment of participants included women who were representative of all self-reported pregnant smokers identified routinely at maternity booking in the National Health Service (NHS) Greater Glasgow and Clyde Health Board area . The prip < 0.0001). The mean birth weight of women who stopped smoking was 3473 g, compared to 3065 g in women who continued to smoke, a difference of 408 g . Given these differences, it might be expected that the difference in mean birth weight between randomised groups would be 0.143 × 408 g, or 59 g. In fact, the difference in birth weight between randomised groups was only 21 g .In the incentives group, 68 women (23.1%) were confirmed non-smokers at 34–38 weeks gestation compared to 25 women (8.7%) in the control group, a difference of 14.3%  analysis, unlike other trials , showed Statistical analyses were performed with R for Windows v3.2.4  and SAS Table Those women in the incentives group who continued to smoke were doing so despite the offer of incentives to quit. In that respect, they can be seen as a group of “hardened smokers”, who could not be encouraged to quit with the incentives on offer. It is not unreasonable to think that these women may lead the least healthy lifestyles of those in the trial. These are the women who had the smallest babies.Similarly, those in the control group who managed to quit, did so despite not being offered any additional incentive, beyond the health benefits of stopping smoking. These women can be viewed as “independent quitters”, i.e. the least entrenched in their habit, and most able to quit. It may well be the case that these women adopted other healthy lifestyle choices during their pregnancies. These women had the largest babies.Following this logic, the control group women who continued to smoke can be seen to be made up of two subgroups: the hardened smokers as aforementioned, plus women who might have been able to quit, if they had been offered incentives. By the same token, women in the incentives group who quit smoking likely includes some independent quitters, plus women who managed to stop smoking as a result of being offered incentives (and who might not have quit had they not been offered any incentives).Therefore, both the control group smokers, and the incentives group quitters, include women in a third subgroup who, when offered incentives are able to stop smoking, but without incentives, continue to smoke. This group could be viewed as “potential quitters”, who probably tend to make (or are more able to make) healthier lifestyle choices than the hardened smokers, but not to the same extent as the independent quitters.HS, PIQ and PPQ, respectively, then from the proportion of the incentives group who continued to smoke, we can estimate PHS as 0.769 (76.9%); from the proportion of the control group who managed to quit, we can estimate PIQ as 0.087 (8.7%). Therefore, we can estimate the proportion of potential quitters, PPQ, as 1 − 0.769 − 0.087 = 0.143 (14.3%). This is exactly the same percentage as the difference between the incentives group and the control group who managed to quit.If we define the probabilities of being a hardened smoker, an independent quitter or a potential quitter as PHS, can be estimated from the mean birth weight of babies born to women in the incentives group who continued to smoke, i.e. 3053 g. The mean birth weight for a baby of an independent quitter, WIQ, can be estimated using the mean birth weight of babies born to women in the control group, who managed to quit, i.e. 3586 g.The mean birth weight for a baby of a hardened smoker, WPQQ), or who are not offered incentives and therefore continue to smoke (WPQS). The difference between these two values, WPQQ − WPQS, can be seen as the gain in birth weight attributable to the offer of incentives, for babies born to women who are potential quitters. However, we do not have any direct means of estimating these two mean birth weights.What we are interested in is the mean birth weight of babies born to potential quitters who are either offered incentives and manage to quit (W1) can be thought of as a weighted mean of the birth weight of babies born to the independent quitters (WIQ) and the potential quitters who were able to quit (WPQQ) within this group:The mean birth weight of babies born to women in the incentives group who managed to quit (W0) can be seen as a weighted mean of the birth weight of babies born to the hardened smokers (WHS) and the potential quitters who were not able to quit (WPQS):so thatSo, the difference in mean birth weight attributable to the offer of incentives, amongst those women with the potential to benefit, can be estimated as 3338 − 3193 = 145 g.In order to assess the statistical significance of this difference, we applied a bootstrapping procedure, with 10,000 replicated datasets, giving a 95% confidence interval for this difference of −617 g to +803 g. This confidence interval is wide, and includes zero, so there is no statistical evidence of an increase in birth weight attributable to the intervention.The bootstrapping procedure also provided confidence intervals for the estimates of mean birth weight for each subgroup separately: the hardened smokers, independent quitters, and the potential quitters with incentives (who continued to smoke) or with incentives (who were able to quit), as shown in Fig. 1) and the control group (μ0), are viewed as weighted means of the mean values in these two subgroups of participants:c and πn are the proportions of compliant and non-compliant individuals in the population, μn is the mean response for non-compliant individuals (which is the same in the intervention and control groups), and μc0 and μc1 are the mean outcomes for compliant individuals, which depends on whether the individual is randomised to the intervention or control group (and therefore receives the intervention or not). Under this formulation, the CACE can be estimated as:In the simplest formulation of a CACE analysis, participants are classified as compliers or non-compliers with the intervention. For a continuous outcome, the mean value in the intervention group . However, the general CACE formulation begins to make sense when we see that it is an instrumental variable analysis: in general, the CACE analysis treats randomisation to the intervention as an instrumental variable, which can only affect the outcome via the mediator (receiving the intervention). The language used in the CACE formulation maybe does not fit, but the principles do: in our data, we assume that the instrumental variable of randomisation to receive the offer of incentives to stop smoking can only affect birth weight via the mediator of quitting smoking. “Compliers” are now seen as women who quit when randomised to be offered incentives, but continue to smoke when randomised to not receive the offer of incentives, “non-compliers” are those who continue to smoke whether randomised to be offered incentives or not, and “always takers” are women who quit regardless of which group they are randomised into. We can assume that there are no “defiers” .The above description of the CACE analysis is actually a simplification of a more general formulation. In general, the population can be divided into four groups: compliers (who receive the intervention if it is offered), non-compliers (who do not receive the intervention when it is offered), “always takers” (who receive the intervention whether it is offered or not), and “defiers” . The simple formulation described above is based on the (not unreasonable) assumption that in a randomised trial, there is no opportunity to be an always taker or a defier , i.e.:Therefore, the CACE analysis simplifies to the average birth weight of babies of intervention participants (3141 g) minus the average birth weight of control participants (3120 g) divided by the proportion of women who stop smoking only when offered an incentive. Since the total proportion who quit smoking in the incentives group is πThis is exactly the same value as was obtained by the intuitive approach described earlier. In fact, the two approaches are the same; the group we identified as “independent quitters” are the “always takers” of CACE notation, the “hardened smokers” are “non-compliers”, and “potential quitters” are “compliers”.The estimated increase in birth weight caused by the offer of incentives in the CPIT trial was 21 g overall, but for the subgroup of women who quit as a result of the offer of incentives, and would not have been able to do so without the offer, it was 145 g. Throughout, we have viewed this as the causal effect of the intervention, within the subgroup who are susceptible to (comply with) it, but the value of 145 g can equally be seen as the causal effect of stopping smoking on birth weight, within this subgroup of women for whom the offer of incentives is sufficient to help them to quit. This is in line with birth weight differences associated with smoking cessation recorded by other researchers - 160 grams for light smoking (1–9 cigarettes per day) and 230 grams for heavy smoking (>9 cigarettes per day) .To try to understand this analysis, Fig. Figure We initially thought that a CACE analysis was not relevant, since the notion of compliance and non-compliance  was not applicable in a trial of the offer of incentives, which was received by all women randomised to receive it (and none of those randomised to not receive it). We therefore pursued an intuitive analysis, estimating the mean birth weight of two subgroups of women unaffected by the intervention (hardened smokers and independent quitters), leading to estimates of the mean birth weight for a third group of potential quitters, with and without the offer, and therefore to the estimate of the causal effect of the intervention for these women.We found, however, that the CACE approach could be used, even though the terminology was slightly incongruous in this setting. This gave us some reassurance that our intuitive approach was valid. Also, whilst the CACE approach gave an estimate of the causal effect of the intervention, the intuitive approach provided us with estimates of the mean birth weight of each subgroup separately, and for the potential quitters, of the mean birth weight with and without the offer of incentives. We feel that this has given us a greater understanding of the characteristics of these subgroups and the impact of the intervention.It should be noted that: (1) the trial aimed to examine the effect of the offer of financial incentives in addition to routine smoking cessation service support to engage with smoking cessation services and/or quit smoking during pregnancy. This means that examining the effect on birth weight only in potential quitters (those who can quit only with the offer of incentives) may be of limited public health importance; (2) neither of these estimates is anywhere near statistically significant (95% CI −615 g to +803 g). In order for a trial to have 80% power to detect an effect on birth weight, if we take 100 g as a clinically important effect (in those who manage to quit as a result of the intervention), then we estimate a study of 27,637 participants  would be required. This assumes that 14.3% of women stop smoking as a result of the offer of incentives, so the mean difference in birth weight between randomised groups will be 14.3 g. This is clearly not feasible. In order to improve the power of a trial to detect an effect on birth weight, then the intervention itself must be improved to increase the impact on quit rates (e.g. by optimising the package of incentives on offer), the intervention must be targeted at those most likely to be a potential quitter , or the analyses outlined in this report must be improved to estimate the causal effect of the intervention on birth weight.Since completing the analyses described in this report, we have become aware of two avenues for further work. Our analyses are the same as a very simple two-stage least squares analysis ; we confNevertheless, whilst an overall 21-g average improvement in birth weight may appear trivial, our analyses show that for individual mothers who benefit from the intervention and manage to give up smoking during pregnancy as a result, there may be a substantial increase in the birth weight of their infant.This study uses secondary outcome data from a randomised controlled trial. It adds to evidence from trial data on smoking cessation in pregnancy in relation to improvements in birth weight . Our anaThis analysis supports the offer of financial incentives to help pregnant smokers to stop smoking during pregnancy in order to improve outcomes for infants, by increasing their birth weight.In the current financial climate, advocating the offer of financial incentives to pregnant smokers to engage with smoking cessation services and or quit smoking is difficult. Demonstrating that the clinically insignificant improvement in average birth weight of 21 g hides a clinically significant estimated increase in birth weight of 145 g, amongst those women who manage to give up smoking as a result of the intervention, helps to justify the offer of incentives for smoking cessation as a reasonable intervention to consider.Additional file 1:Flow diagram of the CPIT II . (DOCX 2Additional file 2:Consort checklist submitted with the manuscript for the CPIT II . (DOC 21Additional file 3: Table S2.from the CPIT  (DOCX 13


The purpose of this study is to examine the relationship between infections and birth outcomes in pregnant Chinese women by using propensity score (PS) matching. The data used here was from a large population-based cross-sectional epidemiological survey on birth defects in Shaanxi province, Northwest China. The babies born during 2010–2013 and their mothers were selected with a stratified multistage sampling method. We used PS-matched (1:1) analysis to match participants with infections to participants without infections. Of 22916 rural participants, the overall prevalence of infection was about 39.96%. 5381 pairs were matched. We observed increased risks of birth defects with infections, respiratory infections and genitourinary infections during the pregnancy . There was also a significant increase of low birth weight associated with respiratory infections (1.13(1.01–1.27)). The association of birth defect with the infection could be relatively stable but the effect could be mediated by some important factors such as mother’s age, education level and economic level. The infection during pregnancy is common in Chinese women and might increase the risk of offspring birth defects and low birth weight, especially in younger, lower education, poor pregnant women. As a widely existing health threat, adverse birth outcomes such as birth defects, low birth weight (LBW) and preterm birth (PTB) can produce great disease burden on fetal health status at birth, as well as potential developmental problems in childhood and future the risk of various diseases4. For example, birth defects, also known as the congenital disorder, are a condition existing at or before birth regardless of cause. They are the primary causes of infant mortality, and different disabilities are in developed countries. Congenital anomalies resulted in about 632,000 deaths per year in 2013 down from 751,000 in 19905. Birth defects are present in about 3% of newborns in the USA6. In China, CHD is the most prevalent birth defect, accounting for over 10% of cases of infant mortality in 20087. As for another adverse birth outcome, LBW (<2500 g) also contributes a lot to infant mortality and childhood morbidity, as well as individual health during the life course8. More than 20 million infants worldwide, accounting for 15.5% of all births, are born with LBW, and 95.6% of these babies occurred in developing countries9, especially in rural areas. Preterm birth (<37 weeks of gestation) is the main cause of neonatal death and the second leading cause (after pneumonia) in children under 5 years of age10. The estimated number of PTB infants was approximately 15 million in 2010 worldwide, which represented 11% of all live births11.Maternal and child health is a part of the most important global public health problems, which has been included as a notable part of United Nations Millennium Development Goals12. For instance, lack of maternal multivitamin use13, maternal age, education and economic level, maternal use of drug13, alcohol14, and cigarettes15 during pregnancy have all been associated with the occurrence of fetal anomalies.The mechanisms of adverse birth outcomes are incompletely understood. Throughout gestation, they are generally considered induced by the complex effects of genetic and multiple environmental risk factors18. Infections are the result of infectious agents including viruses, viroids, prions, bacteria, nematodes and so on. For instance, pregnant women are vulnerable to viral respiratory infections19 and are found to be at high risk for influenza and related complications20. Influenza during pregnancy has been linked to spontaneous abortion, stillbirth, and prematurity22. Maternal fever is also common during pregnancy. Botto et al. estimated that 5–10% of women, on average, report a fever in early pregnancy23. It was reported that fever has been suspected to harm the developing fetus24. Even short exposures to elevated maternal body temperature have been reported to lead to cell death, vascular disruptions, and placental infarction25. In addition, urinary tract infections, especially pyelonephritis26 are reported to associate with preterm birth.There is also growing evidence that some infections during pregnancy may play an important role in the occurrence of adverse pregnancy outcomes and congenital anomaliesHowever, these studies have a few limitations. Prospective cohort studies with laboratory confirmation of infections are ideal but are often hampered by low numbers of events, particularly for outcomes such as fetal birth defects. Existing birth registries can facilitate large studies, but they do not routinely collect other more confounders. In other words, a number of study design issues have limited the interpretation of previous results because of small sample sizes and lack of assessment of potential confounders. Moreover, individuals with mild or subclinical infections may never seek medical care. There will have a lower infection rate.27. The propensity score (PS) technique has turned out to be associated with the largest reduction selection bias and residual biases assessing causal effects in observational studies29. PS matching estimates the probability of individuals being “exposed” to infections and allows comparison with “unexposed” people with similar demographic and socioeconomic characteristics, that is, with a similar profile of potential confounders. Using a PS approach to estimate differences in adverse birth outcomes between groups of people with and without infections is an important alternative method to “standard” multivariable regression adjustment to estimate the unbiased association of infections with birth outcomes, and hence the causal association in the absence of confounding, random error, measurement error, and selection biases. In addition, in sparsely populated Northwestern China, economic condition is worse and health services are less developed than in Eastern and Coastal areas of China. Consequently, this large population-based observational study has been carried out to better understand, by using PS methods, the potential effects of exposure to maternal infections on fetal development and infant health in rural northwestern China. This analysis will be used as a baseline for better quantification and qualification of these risks so as to provide not only research insights into mechanisms of teratogenicity but could be helpful in clinical counseling of exposed women and primary prevention.An observational study, in contrast, is often limited by potential residual biases from measured confounders and possible biases due to unmeasured confounders22916 participants were available for analysis. The overall prevalence of infections during the pregnancy was 39.96% (9157/22916). Of these, the prevalence of respiratory infections and genitourinary infections were 38.99% and 1.82%, respectively. After propensity score matching, 5381 participants with infection were matched one to one to participants without infection. The overall prevalence of adverse birth outcomes was 3444 participants (15.41%). Of these, birth defects were 486 participants (2.12%), LBW was 2773 participants (12.41%) and PTB was 613 participants (2.68%).Before PS-matching, participants’ characteristics varied considerably among the two groups Table . We founTable Table In the present study, there were relatively high prevalences of adverse birth outcomes and high risk of being infection during pregnancy. Our findings demonstrated that the infection during the pregnancy was associated with increased risk of birth defects and low birth weight. There was also an increased risk of infections and preterm birth but remained not significant statistically. In addition, the association of birth defect with the infection could be stable relatively but the effect of the infection could be mediated by some important factors such as mother’s age, education level, economic level and gender of the baby. These findings are important to strengthen the guidance of pregnancy and reduce the incidence of adverse birth outcomes.31. Shahrukh et al. observed a modestly increased risk  evaluating the association between maternal fever and Oral Clefts13. Botto et al. also found febrile genitourinary infections were associated with selected heart defects, particularly right-sided obstructive defects  and possibly others23.Our finding was supported by several studies that found significant associations between congenital defects and febrile illness caused by infectionset al. explored an association between birth weights with the maternal respiratory infection during pregnancy in the community settings32, which was similar to our study. Philpott et al. also found a significant effect of febrile illness during pregnancy on LBW33.Low birth weight is the result of either preterm birth or poor growth of the fetus resulting in intrauterine growth restriction. After PS-matching, the risk of LBW with respiratory infections increased to 1.13(95% CI: 1.01–1.27) from 1.12(95% CI: 1.01–1.24). Asad 35. However, there were similar results from other researches, which have not shown an association with preterm birth37. Erin et al. also did not find a significant effect of febrile illness during pregnancy on PTBs33. These differences may be the result of inaccuracies in maternal self-report of an infection, which do not always provide with a specific definition. The prevalence of infection could be underestimated or overestimated. In addition, the differences between the region and the population should also be accounted for it.However, we did not observe any strong differences in associations between infections and PTB. This finding contrasted with the general opinion that maternal febrile infections arising during the preterm period can trigger the onset of laboret al.13 found an increased risk of oral clefts in women who reported a common cold or influenza and did not take drugs during the illness. In comparison, women who took the medications had no increased risk, which was in complete agreement with our subgroup analysis. Furthermore, hormones, such as progesterone and testosterone, may have regulatory roles in renal development and could potentially influence renal outcomes by gender30. Therefore, it is also necessary that the prevention of birth defects should be focused on these pregnant women.We further conducted subgroup analysis and found that the association of birth defect with the infection could be stable relatively. However, the effect of the infection could be mediated by some important factors such as mother’s age, education level, economic level, taking medicine during the pregnancy and gender of the baby. Similar to our findings, Shahrukh 38. Activation of increased body temperature is associated with interruption of protein synthesis and enzyme production, which results in cellular processes  becoming altered or dysfunctional38. Therefore, one mechanism by which maternal hyperthermia might influence fetal development is unregulated apoptosis. Furthermore, it has been suggested that maternal exposure to hyperthermia induces the transcription of protective heat shock proteins, a process that might potentially divert resources from normal protein synthesis away from normal development39.A mechanistically plausible explanation of our findings is that when the infection in pregnancy occurs, the maternal immune system is mobilized, causing changes in the level of cytokines in the fetal environment. Some cytokines, such as tumor necrosis factor-a, interleukin 1 and interleukin 6 are pyrogenic, causing hyperthermia to occur through alteration of the set point in the hypothalamus29. After PS-matching, all selected potential confounding factors were sufficiently balanced between the two matched groups in our study. This is analogous to randomization procedures used in clinical trials, as on average the distribution of covariates will be balanced between the infection and non-infection groups, which strengthens causal inference and thus improves the methodological quality of observational research. Compared with conventional matching, PS matching can consider more matching factors and improve the research efficiency. Several limitations should also be addressed while interpreting our results. Firstly, any cause-effect conclusion cannot be drawn due to the limitations of the cross-sectional design, although we used PS-matching to control biases as much as possible. Secondly, the diagnosis of infection may be imprecise due to the self-report and the potential mis-classification of maternal infection occurred. However, the pregnancy is one of important life events for the women and they could care about many events related to pregnancy such as infection during pregnancy. But it might still underestimate the associated effect of infections during pregnancy and adverse birth outcomes because of recall bias. Unfortunately, it is difficult to get laboratory evidence or clinical diagnosis in a large-sample cross-sectional study. Thirdly, we selected whole pregnancy as exposure period rather than a definite stage due to the low incidence of adverse birth defects. Future studies should attempt to consider this effect of differential critical periods using different research design such as case-control study.This study is the first and largest survey that has presently been conducted in Northwest China, and provides up-to-date data on birth defects. In addition, PS-matching method was used to study the relationship between infections and birth defects, LBW, PTB in China, which can reduce confounding as observed covariates are balanced at each particular value of the propensity scoreOur findings suggest that the infection during pregnancy in Chinese women was associated with increased risk of birth defects and low birth weight, especially in younger, lower education, poor pregnant women. Therefore, the effective infection control measures in antenatal care should be enhanced further and the high-risk female population should be focused in order to prevent and control adverse birth outcomes.40 and the ethics committee of the Xi’an Jiaotong University Health Science Center approved the study (No. 20120008). All research was performed in accordance with relevant guidelines and regulations and informed consent was obtained from all participants and/or their legal guardians. The datasets analysed during the current study are available from the corresponding author on reasonable request.The data used here was from a large population-based cross-sectional epidemiological survey on birth defects in Shaanxi province, Northwest China between August and December in 2013. The babies born during 2010–2013 and their mothers in this survey were selected with a stratified multistage sampling method. Comprehensive methods of this survey have been previously published elsewhereAll family-level primary data was reported by the mothers in the pre-coded structured family questionnaire, including socio-demographical information, information regarding access to maternal healthcare, health status, and lifestyles during pregnancy, and child-level information such as birth defects and another birth data. Information on birth defects was also collected in the pre-coded structured birth defects questionnaire which contained questions on birth outcomes, diagnostic information at local hospitals, time of diagnosis and the types of defects. Medical records at local hospitals were used as final diagnosis references. In the survey, all questionnaires were designed by Xi’an Jiaotong University Health Science Center. Only after obtaining written informed consent, face-to-face interviews were carried out by the field investigators.Taking into account the higher percentage of adverse birth outcomes in rural areas, in our present study, we only analyzed the relationship between pregnancy infections and adverse birth outcomes in rural participants. Therefore, the inclusion criterion was a single live birth in rural areas in this article. Excluded were cases of abortion, stillbirth, multiple births, or neonates without the information of having an infection or not. Finally, 22916 singleton live-birth newborns enrolled in this study.4. Therefore, adverse birth outcomes in this study included these three outcomes. Birth defects were recorded as ICD-10 codes and were defined as cleaned malformations, i.e., excluding previously defined insignificant diagnoses41. LBW refers to birth weight less than 2500 g and PTB refers to birth at less than 37 weeks gestation. These above outcomes required the definite diagnosis of the hospital.Previous studies have shown that birth defects, low birth weight (LBW) and preterm birth (PTB) play an impartment role in producing disease burden on fetal health status at birth, as well as potential developmental problems in childhood and future the risk of various diseasesData on maternal infections during entire pregnancy were gathered from mother’s self-report or through treatment records if they sought medical care. The reported infections were divided into three types: respiratory infections, genitourinary infections and other infections. Respiratory infections included influenza/common cold , bronchial infections and pneumonia; Genitourinary infections included infections of the kidney, bladder, urinary tract and genital tract; Other infections included gastrointestinal infections and some infections affecting multiple organ systems  during pregnancy. In brief, maternal infection episode was defined as having any one of the above infections during entire pregnancy.30, such factors as some socio-demographic characteristics, lifestyle, folic acid supplement, taking medicine and family history of birth defects were found associated with adverse birth. So these factors could confound the association between infection and adverse birth outcomes. In our analysis, these factors were controlled. Socio-demographic characteristics captured by the survey included maternal age (<30 and ≥30 years), education , socioeconomic status  and baby gender . The socioeconomic status was measured using the Demographic and Health Survey wealth index42. The principal component analysis was utilized to combine the variables representing families’economic level. On the basis of the first principal component, then, the socioeconomic status of the families was classified into the tertiles representing poor, medium and rich families. Maternal passive smoking and drinking during pregnancy had two dichotomous outcomes variables (no and yes). The folic acid supplement was defined as the number of times when the respondent took folic acid greater than 30 days, and similarly, also classified into two categories (no and yes).Based on previous studiesIn the study, the categorical variables were expressed as frequency and percentages and were compared using chi-square tests between the two independent groups. Ordinal data such as education and wealth index were compared using Rank-Sum Test. Logistic regression was performed to investigate the association between infections and birth outcomes. With the purpose of comparing with PS-matching, multivariate logistic regression was also performed before PS-matching. The model included all kinds of birth outcomes as the dependent variable, infections as the exposure, and all the variables mentioned above as potential covariates.29. We performed 1:1 matching by using the calipers matching, which matches each person with and without an infection who has similar propensity scores, within a caliper of 0.000143. The variables selected for the propensity score model included all the covariates above. To evaluate the success of balancing the baseline characteristics between the matched groups, standardized differences were estimated, which directly quantifies the bias in the proportions of covariates across the groups, expressed as a percentage of the pooled standard deviation (SD)45. The standardized difference was calculated as follows:To minimize the possible bias arising from the study design, we constructed a PS model to adjust for potential confounding variables for having an infection or notP1 and P2 represented the prevalence of confounders observed in the exposed and unexposed groups, the percentage standardized difference is:If whereAfter PS-matching, univariate logistic regressions were performed to explore the association of infections with adverse birth outcomes. Subgroup analyses were conducted further based on selected covariates with the purpose of assessing the robustness of our findings regarding the effect of infections on birth defects and initially identifying high-risk groups.Data was entered into Epidata 3.1 by double entry  and all statistical analysis was conducted with STATA version 12.0 software . All statistical tests were evaluated using two-tailed 95% confidence intervals (CI). Statistical significance was set up if P < 0.05.


We found that mean birth weight declined by 1.07 grams/year from 2001 to 2015. After adjustment for gestational length, the decline in birth weight was attenuated (0.37 grams/year). The incidence of both SGA and LGA significantly decreased during the study period . We found a narrowing of disparities in SGA and LGA incidence across different maternal educational levels and residence location. Our results demonstrate that there has been an increase in the proportion of neonates born in the healthy birth weight range in Guangzhou.Estimates of trends in birth weight may be useful in evaluating population health. We aimed to determine whether temporal changes in birth weight have occurred amongst 2.3 million neonates born in Guangzhou, China, during 2001–2015 and investigate the socioeconomic determinants of any changes. Trends in mean birth weight and annualized changes with the associated 95% confidence intervals (CIs) in the incidence of small for gestational age (SGA) and large for gestational age (LGA), defined as birth weight <10 Located in southern China, Guangzhou is one of the country’s most developed cities. Over the past two decades, a comprehensive healthcare system has been implemented, providing near universal health coverage to all citizens. The success of this system is reflected in the decline in maternal mortality rate, which at 7 per 100,000 in 2014, is far lower than the national average2.China has made remarkable progress in improving maternal and child health and reducing maternal mortality, from 64 in 1996 to 22 per 100,000 in 2014 and infant mortality, from 36 in 1996 to 9 per 1000 live births in 20144. Small (SGA) or large (LGA) for gestational age are common indicators of restricted or excessive fetal growth. Previous studies have shown that SGA babies had increased risk of respiratory complications, hypoglycemia and necrotizing enterocolitis during perinatal period, neurological impairment in childhood and cardiovascular disease and type 2 diabetes mellitus in adulthood5, while LGA is associated with increased risk of birth trauma, cesarean section and postpartum hemorrhage as well as obesity, metabolic syndrome and type 2 diabetes mellitus in adulthood6. Therefore, examining trends in incidence of SGA and LGA may be useful in evaluating population health. Conventionally, SGA or LGA have been defined as weight at birth below the 10th or above the 90th centiles when compared to a birth weight reference7. These references were based on the distribution of the weights of all babies born in a specific population, rather than describing healthy or physiological growth7. Recently, a multinational standard for newborn weight was published by the International Fetal and Newborn Growth Consortium for the 21st Century (INTERGROWTH-21st)7. This study demonstrated that in mothers free from social, nutritional, medical and other constraints on fetal growth, the growth of babies around the world is remarkably similar. The INTERGROWTH-21st birth weight standard therefore provides a valid instrument to compare fetal growth and birth weight throughout the world.Secular trends in birth weight can indicate changes in maternal health and nutrition. Babies with birth weights outside the normal range have higher risks of mortality and morbidity in the perinatal period and later in life8, had lower mean birth weight with reduced incidence of SGA and LGA. It is possible however that this difference may not represent the actual long-term trends. We aimed to examine trends in mean birth weight and incidence of SGA and LGA among 2.3 million singleton births during 2001–2015 and whether these trends reflected changes in maternal characteristics including indicators of changing socio-economic status.Data available on long-term trends of birth weight in China have been limited. We previously demonstrated in Guangzhou that babies born in 2012, when compared to 2002Maternal and infant characteristics by 3 discrete 5-year periods  are shown in Table Table Table Figure In one of the most developed cities in China, we found mean birth weight declined by 1.07 grams/year from 2001 to 2015. This decline was largely explained by shorter gestational length. We have demonstrated that during this same time, both SGA and LGA incidence decreased , which cannot be entirely explained by changes in maternal and infant characteristics. By using education and residence location as proxies for maternal socio-economic status, we present evidence that the disparities in SGA and LGA risk between mothers with different socio-economic status has narrowed.15. In the United States, mean birth weight among term singleton neonates decreased from 3441 grams in 1990 to 3389 grams in 2004 (3.0 grams/year), independent of decreasing gestational length12. Such data from China have been relatively sparse and mostly short-term. An analysis from southeast China found the mean birth weight for term and post-term infants decreased from 3378 grams in 2000 to 3369 grams in 200516. By contrast, a recent study in Shaanxi province (northwestern China) reported the rates of low birth weight decreased while those of macrosomia increased, indicating an increase in higher birth weights in this population17. Our data demonstrate that there has been a decrease in mean birth weight in Guangzhou. Gestational age is a strong determinant of birth weight; thus changes in gestational age at birth may result in trends in birth weight. Over the 15-year study period, we observed mean gestational length in Guangzhou decreased from 39.02 to 38.79 weeks (1.6-day difference). After adjustment for gestational length, the downward trend of mean birth weight was largely attenuated, suggesting the shortening of gestation length was likely to be responsible for the observed decrease in mean birth weight, which is in contrast to the observations from the United States12.Studies of secular trends of birth weight in developed countries have demonstrated birth weight has begun to decline during the last three decadesWe observed that the incidence of SGA and LGA in our population decreased over time. After adjusting for maternal age, parity, infant sex and gestational length, the decreasing trend of SGA incidence was attenuated but remained significant, while that of LGA incidence was strengthened, indicating these covariates did not, at least not fully, explain the decline in incidence of SGA and LGA. A plausible explanation of the observed trend is the improvement of maternal care, resulting from a variety of public health programs in Guangzhou that aimed to promote maternal and child health . On the other hand, the decreasing trend of SGA incidence in current analysis is more evident (from 8.0% in 2001 to 5.6% in 2015) than that in our previous report (9.2% in 2002 and 8.6% in 2012). These differences might be due to the fact that in the current analysis we have included a much larger dataset (2.3 million babies) than previous study (<90 thousand babies) and used the INTERGROWTH-21st standard to define SGA and AGA instead of local birth weight reference8. In the current study we also found narrowing disparities in incidence of SGA and LGA across maternal educational levels and residence locations. This might also be attributed by the implementation of above-mentioned public health programs in Guangzhou st standard to define SGA and LGA. To the best of our knowledge, until now this new standard has not been used to inspect temporal trends in birth weight in any population. Some studies have advocated the continued use of local or customized charts19; however these local charts are only relevant to the population and time from which they were derived and hence make comparison between populations and studies impossible.The major strength of present study is the use of a large (over 2.2 million births) and up-to-date (until 2015) dataset covering a sufficiently long period of time (15 years) to examine the trends in birth weight. In addition, we used the INTERGROWTH-2120. Secondly, although it is standard practice to confirm gestational age using ultrasound examination during first or early second trimester in Guangzhou21, the last menstrual period-based method was also used in some rural areas, which may have resulted in miscalculation of gestational age. Lastly, we have no data on length and head circumference at birth, two other important indicators of fetal growth, in this dataset.Some limitations need to be mentioned. Firstly, because the Guangzhou Perinatal Health Care and Delivery Surveillance System (GPHCDSS), from which we obtained the data, was not designed for research purposes, some variables relevant to fetal growth were not collected. This meant that we were unable to examine the contribution of pre-pregnancy body mass index (BMI), gestational weight gain, maternal diet and smokingIn conclusion, although mean birth weight decreased slightly, we observed a substantial improvement in key fetal growth indicators (SGA and LGA) in Guangzhou during 2001–2015. This occurred in parallel to improvements in maternal health care. Birth size is a strong predictor of not only short-term health outcome but also long-term consequences during adulthood. Optimization of birth weight for gestational age could have a significant influence on child health. By sharing the experience learnt in Guangzhou, we hope that this improvement may be replicated as other regions in China and other low- and middle-income countries.23. In brief, the electronic GPHCDSS database was implemented in 2000, and covers over 99% of deliveries in Guangzhou (population: 13 million). Dedicated trained health workers are responsible for collating and registering birth information at each hospital. Validity was confirmed at various levels: by the Chief Midwife, the Chief Physician, and the Department of Medical Administration at each hospital; by the newborns’ parents at the time when the birth certificate was issued; and by the Guangzhou Municipal Health Bureau through annual sampling survey. From this surveillance system information on maternal demographics, pregnancy and delivery was collected.Birth data between 2001 and 2015 were acquired from the Guangzhou Perinatal Health Care and Delivery Surveillance System (GPHCDSS), which has been described elsewherest Birth Weight Standards7. We also obtained information from the GPHCDSS dataset on maternal and newborn characteristics, including maternal age at delivery and educational level, parity, delivery mode, date of birth, newborn’s sex, gestational age and birth weight.This analysis included all singleton live births delivered in Guangzhou from 2001 to 2015, with gestational ages between 33 and 42 completed weeks of gestation that in line with the INTERGROWTH-21A total of 2,388,629 records of live births born in 2001–2015 were retrieved and 97,884 were excluded due to duplicate records (n = 8826), multiple pregnancy , missing or implausible data on birth weight (n = 1274), sex (n = 31), and gestational age (n = 999), gestational age of less than 33 weeks , or more than 42 weeks . As a result 2,290,745 records of singleton live births were included in the analysis. This study was conducted in compliance with local and national regulations and was approved by the institutional ethical committee board of Guangzhou Women and Children’s Medical Center. Data used in this study were anonymous, and no individual identifiable information was available for the analysis.th or above 90th centile of a sex- and completed gestational age specific birth weight, according to the published INTERGROWTH-21st standard7. Infants having birth weight between 10th and 90th centiles were considered to be AGA.Birth weight (in grams) was measured routinely by registered midwives using an electronic weighing scale within half an hour of delivery. Gestational age at birth was determined based on an ultrasound examination in the first or second trimester and was expressed as completed weeks. Where ultrasound data were unavailable, the last menstrual period was used to calculate gestational age. We defined SGA or LGA as birth weight below the 10Descriptive statistics of maternal and infant characteristics are shown in 3 discrete 5-year periods , including maternal age , educational level , parity (1 and ≥2), residence location , delivery mode  and infant sex . We then calculated the mean values of birth weight, gestational length, the incidence of SGA and LGA for each year between 2001 and 2015. Trends in birth weight and gestational length assessed using linear regression models, with birth year included as a continuous variable, and trends in SGA and LGA were examined using Poisson regression models with robust variance to estimate risk ratios (RRs) and 95% confidence intervals (CIs) for the change per year. We presented 3 models to explore the trends in outcomes: Model 1 was unadjusted to quantify the crude trend; Model 2 was adjusted for maternal age, parity and infant sex; Model 3 was further adjusted for gestational length that had a decreased trend. If changes in covariates accounted for the observed trends in the outcomes, adjustment for these variables would attenuate the estimated RRs of birth year. We did not adjust for maternal education in the model because data on education level had not been routinely collected since 2014 and the estimates of RR changed only slightly when included education level in the model.We then presented the trends in incidence of SGA and LGA graphically, stratifying by maternal age groups, educational level and residence location, respectively. To describe the difference in risk of SGA and LGA among these subgroups, we calculated RRs and 95% CIs of each subgroup for each year separately. We also included multiplicative interaction terms between birth year and maternal age groups, educational level, residence location in Model 3 separately to test whether the difference in risk of SGA and LGA among these subgroups differed over the 15-year period.We used SAS version 9.2  for all statistical analyses.The data that support the findings of this study are available from Guangzhou Women and Children’s Health Information Center but restrictions apply to the availability of these data, which were used under license for the current study, and so are not publicly available. Data are available from the authors upon reasonable request and with permission of Guangzhou Women and Children’s Health Information Center (contact Xiao-Yan Xia at xiaxy0656@163.com).Supplementary Information


Despite evidence that genetic factors contribute to gestational length and pretermbirth, robust associations with genetic variants have not been identified. Wehypothesized that analyzing larger data sets with gestational length information bygenomewide association would reveal trait-influencing variants.P < 5×10-8) or suggestiveassociation (P < 1×10-6).We performed a genomewide association study in a discovery data set of 43,568 women ofEuropean ancestry from 23andMe, Inc., for gestational length as a continuous trait andfor term or preterm (<37 weeks) birth as a dichotomous outcome. We used three Nordicdata sets  for replication of 14 genomic loci achieving either genomewide achieved genomewidesignificance, all of which were replicated in the Nordic data sets. Functional analysisof the WNT4 locus indicated the likely causative variantalters the binding of ESR1. ADCY5 and RAP2C, which had suggestive significance in the discovery stage, weresignificantly replicated and achieved genomewide significance in joint analysis. Commonvariants in EBF1, EEFSEC andAGTR2 were also associated with preterm birth withgenomewide significance. Analysis of mother-infant dyads indicated that these findingslikely resulted from maternal genome actions.In the discovery stage, for gestational length, four loci (Our study is the first to identify maternal genetic variants robustly associated withgestational length and preterm birth. Roles of these loci in uterine development,maternal nutrition, and vascular control support their mechanistic involvement andcreate opportunities to investigate new risk factors for prevention of pretermbirth. Preterm birth (defined as birth before 37 completed weeks of gestation) affects 9.6% ofpregnancies in the United StatesPreterm birth, and gestational length in general, is a complicated phenotype withcontributions from two genomes – maternal and fetal – that may have separateor interacting contributions. Furthermore, different genotypes may predispose to pretermbirth at different gestational ages. Finally, defining preterm birth as a dichotomous traitbased upon a somewhat arbitrary cutoff of 37 weeks, rather than time of birth for aspecified level of fetal maturity or as a continuous trait, limits data interpretation.Therefore, defining the genetic variants associated with gestational length (a quantitativetrait) as well as preterm birth (a dichotomous trait), will both yield important newinsights. Further, analyzing gestational length as a continuous trait increases the power todetect associations that is limited when traits are dichotomized.P <1×10-6) and performed replication analyses for gestational length andpreterm birth in three data sets of Nordic women . Further, we provide evidenceindicating the observed effect was due to an action in the maternal genome and providefunctional data implicating the causative SNP underlying the WNT4 locus.To overcome previous sample size limitations, we leverage data on gestational length andpreterm birth in a large sample of women of European ancestry collected as part of genotyping and phenotyping efforts by 23andMe, Inc., a geneticscompany. We then selected the top loci .http://www.eandireview.com).Unrelated women of European ancestry who self-reported gestational length of their firstlive singleton birth were included in the analysis. Categories of preterm birth addressedon the survey were 1) spontaneous preterm labor, 2) planned or required delivery formedical reasons, 3) cervical problems, 4) other, or 5) none of the above. Women with amedical indication for their preterm delivery were excluded from the study; those that didnot specify a medical indication on the survey were retained to optimize sample size.Preterm birth status was determined based on dichotomization of self-reported gestationallength . For those in the preterm group,96.8% of women responded to the question regarding mode of delivery. For the term birthgroup, we ascertained information on aggregate outcomes of all pregnancies, and could notunambiguously determine spontaneous or medically indicated birth at more than 37weeks.Women in the discovery data set were participants in 23andMe’s research program.All women provided informed consent and answered surveys online following a human subjectsprotocol, reviewed and approved by Ethical & Independent Review Services, a privateinstitutional review board . The SNP with smallest P value withineach region was chosen as the index SNP. Regions that achieved suggestive significance(P < 1×10-6) were tested in thereplication stage.We clustered SNPs into association regions (or loci). Specifically, we definedassociation regions by first identifying SNPs with We used the data of 8,643 mothers from three independent Nordic birth studies  of singlGenotyping of the Nordic studies was conducted using various SNP arrays as previouslydescribed.P values (inflation adjusted) combining resultsfrom the three Nordic data sets were calculated using the fixed-effects inverse-variancemethod. Significant replication P values and the samedirection of effect at the index or other significant SNPs  in the region or theirclose proxies (r2 > 0.8) were regarded as statistical evidence of replication of a putativelocus. The significance level of each region was corrected by the effective number ofindependent SNPs tested in the region and the total number of regions that underwentreplication attempts  were also performed to examine possible dominance effect. Thereplication attempts . A regioWe also performed association tests in 4,090 infant samples and joint maternal/fetalgenetic association analysis in 3,184  mother/infantpairs from the Nordic data sets to evaluate whether the observed significant associationswere likely to be of maternal or fetal origin.We checked whether the SNPs associated with gestational length or preterm overlap withpreviously reported genomewide association SNPs in the GWAS catalogWNT4locus, one of the most significant loci with plausible functional relevance in pregnancy.First we examined the expression level of WNT4 in humanendometrial stromal cells, before and after decidualization using mRNA-seq technology. Wepredicted specific transcription factors binding using a Bioinformatic approach andstudied the presence of H3K4me3 marks and open chromatin domains overlapping thehypothetical causal SNP by ChIP-seq and ATAC-seq, respectively. We performedelectrophoretic mobility shift assays (EMSA) to determine whether the variantdifferentially affected specific transcription factor binding. Detailed description ofthese analyses can be found in the WNT4 locus).We performed experimental functional follow-up of the P =2.3×10-41) with older mothers having shorter gestational length. Of these, four had an association P< 5×10-8  shared associationloci for preterm birth risk. Altogether, 14 independent loci were taken forward forreplication. To confirm the robustness of the association signals, we conducted similarassociation tests in a subset of discovery subjects who explicitly checked“spontaneous delivery” in the questionnaire  and the results weresimilar to those obtained from the full discovery data sets  at the index SNPand other SNPs with P < 1×10-6 in thediscovery stage, and their close proxies (r2 > 0.8). Six loci . In addition, associations ofthe BOLA3 locus with gestational length, and the TEKT3 and the TGFB1 loci withpreterm birth showed marginal significance (P < 0.05). Atthe EBF1, EEFSEC, AGTR2, WNT4 and RAP2C loci, the most significant SNPs in the replication stage were eithersame as or in substantial LD (r2 > 0.6) with the most significant SNPs in discovery stage. However, at theADCY5 locus, the LD between the most significant SNPs inreplication stage and the discovery stage is less substantial (r2 < 0.4). SNPs at the EEFSEC locus showednominally significant dominant effects (P.dom < 0.05). r2 >0.8) with SNPs significantly associated with gestational length orpreterm birth in the discovery stage. Within these loci, there are SNPs reportedlyassociated with complex traits  have also been associated with increased risk of prostatecancer (rs10934853-A)WNT4 locus werepreviously associated with endometriosis,cis-expressionQTLs) based on GTEx dataA number of SNPs with potentially functional impact  are encompassed by the loci we identified as potentially important . HoweverADCY5 locus have been reported to be associatedwith birth weightADCY5, WNT4 and EBF1 loci that are associated with birth weight.ADCY5 and WNT4 loci appear to influence birth weight through the fetal genome and noneof them were in close LD with the SNPs showing significant association with gestationallength; while the SNP (rs7729301) at the EBF1 locus seems toinfluence birth weight through the maternal effect, and the allele (G) associated withreduced birth weight was also associated with shorter gestational length  prior todecidualization in vitro to 29.5 TPM after decidualization in samples run in duplicatefrom two different endometrial stromal cell lines. The genetic loci we identified fall in noncoding regions of the genome, suggesting thatthey will affect gene regulation rather than protein function. To dissect the consequencesof these variants, knowledge of cell-type context in which they are active is essential.The lization. UsingRWNT4 that might be altered by variants associated withgestational length. To this end, we used the CisBP web serverWNT4 locus. These analyses indicate that rs3820282 (r2=0.94 with the index SNP rs56318008), which is located in the first intron ofWNT4 is capable of altering the binding of the estrogenreceptor (ESR1). Specifically, the underlying quantitative data from protein bindingmicroarray (PBM) assaysWNT4 locus is modulation (viars3820282) of the binding of ESR1. rs3820282 is also strongly associated with epithelialovarian cancerWe next sought to identify particular regulatory mechanisms controlling the expression ofOur genomewide association study is the first to identify human genetic polymorphisms thatare significantly and reproducibly associated with gestational length and preterm birth, thesingle greatest contributor to mortality in children younger than five years and a commonsource of morbidity throughout life for those who survive. Our approach demonstrates theutility of using data collected as part of direct-to-consumer genotyping and phenotyping torapidly assemble large data sets capable of revealing contributing loci in particularlycomplex phenotypes such as preterm birth where both maternal and fetal genomes, and manygenes, are likely to contribute to the outcome. By combining the power of a large43,568-person discovery data set and stringent replication by the well phenotyped Nordicdata sets, we identified and replicated six maternal genomic loci robustly associated withgestational length and three of them also associated with preterm birth with genomewidesignificance (P<5E-8) in the joint analysis.EBF1 (early B-cell factor 1),also achieving genomewide significance for preterm birth, has been demonstrated to beessential for normal B cell development,EBF1 confers its effect on birth timingthrough pregnancy-specific mechanisms, or by contributing to more general cardiovascular ormetabolic traits that influence gestation remains to be determined. In addition, theassociation between this locus and gestational length may explain the effect of this locuson birth weight reported by Horikoshi et al.The top four replicating genomewide significant SNPs for gestational length are inbiologically plausible genes. EEFSEC , also genomewide significant for both gestational length and preterm birthrisk, participates in the incorporation of selenocysteine into selenoproteins.Selenoproteins, such as the glutathione peroxidases and thioredoxin reductases, servecritical cellular homeostatic functions in maintaining redox status and antioxidantdefenses, as well as modulating inflammatory responses.EEFSEC are in highLD with SNPs that have previously been associated with age of onset of menarche, expressionquantitative trait loci (eQTLs) for EEFSEC abundance, risk ofprostate cancerAGTR2 , the coding genenearest to a group of X chromosome SNPs achieving genomewide significance in the gestationallength analysis, had suggestive association with preterm birth discovery stage, andgenomewide significance for preterm birth in the joint analysis with robust association inthe Nordic replication. AGTR2 has been suggested to play arole in modulating uteroplacental circulation, and harbors variants that may contribute tothe risk of preeclampsia.EEFSEC, is a potential shared mechanism for thesegenes in spontaneous preterm birth.WNT4 (wingless-type MMTV integrationsite family member 4), with strong replication in the Nordic populations. WNT4 mutations have been found in women with Mullerian ductabnormalities, primary amenorrhea, and hyperandrogenism,WNT4, in high LD with our indexSNPs, are associated with risk for endometriosisWNT4, and the T allele generates a strong ESR1binding site, and as such likely alters estrogen-based regulation of WNT4 and/or adjacent genes. The role of estrogen signaling as the functionalconsequence of the polymorphism is further supported by the association of the same regionwith endometriosis and ovarian cancer, both hormone-responsive disorders. Further, theparallel of the spectrum of disorders associated with the WNT locus mirrors that of ARID1A,also critical for endometrial function early in pregnancy, with loss of function variantscausing atypical endometriosis and ovarian cancer, and enhanced estrogen activity. WNT4 did not achieve genomewide significance or suggestiveassociation in the preterm birth risk dichotomous trait analysis, suggesting its role may belargely exerted near term gestation. The final gene locus achieving genomewide significance in the discovery stage forgestational length was ADCY5 (adenylyl cyclase type 5) and RAP2C (member of the RAS oncogene family) achieved near genomewide significancein the discovery stage and were successfully replicated (ADCY5 locus have been reportedto be associated with birth weightRAP2C region (in geneFRMD7) was previously reported to be associated withspontaneous preterm delivery in Danish/Norwegian studies (the samples used in this previousstudy overlap with our replication samples).BOLA3, TEKT3 and TGFB1), while showing marginal evidenceof replication, remain suggestive and await the addition of further studies for analysis. plicated . SNPs atThe primary limitation of our study centers on the characteristics of our study data sets.The gestational length information of the 23andMe samples was self-reported, and 3.2% ofwomen in the preterm group did not respond as to whether the labor and delivery wasspontaneous or medically indicated. In the term group, we were not able to unambiguouslydetermine spontaneous from medically-indicated births. Despite these limitations, weincluded these samples in order to dramatically increase the sample size of the discoverystage, recognizing that our replication data sets would be more precisely phenotyped forspontaneous preterm birth. A previous study suggested that approximately 90% ofmother-reported gestational lengths agreed with their associated medical records.Our study demonstrates the utility of combining large samples with self-reportedphenotyping with more modestly sized but precisely phenotyped replication studies to revealmaternal loci associated with gestational length and preterm birth. With this foundation,future expansion of maternal and fetal genotyped samples associated with gestational lengthinformation is anticipated to further refine our understanding of human pregnancy, risk foradverse pregnancy outcomes, and targeting of new preventive strategies for preterm birth. Asthe National Institutes of Health expand “Precision Medicine” initiatives inthe years ahead, we would argue that the optimal time to advance human health is before andduring pregnancy. Our work suggests that integration of genomic information on women, andlikely their offspring, with birth timing, may allow development of new options forpreventative and therapeutic measures.Supplementary Appendix


Perfluorocarboxylic acids (PFCAs) are environmentally and biologically persistent synthetic chemicals. PFCAs include perfluorooctanoic acid  and long-chain PFCAs (C9–C20). Studies examining long-chain PFCAs and fetal and postnatal growth are limited.We investigated the associations of prenatal exposure to long-chain PFCAs with fetal and postnatal growth.For 223 Taiwanese mothers and their term infants, we measured PFOA and four long-chain PFCAs (ng/mL) in third-trimester maternal serum; infant weight (kg), length and head circumference (cm) at birth; and childhood weight and height at approximately 2, 5, 8, and 11 years of age. For each sex, we used multivariable linear regression to examine associations between ln-transformed prenatal PFCAs and continuous infant measures, and logistic regression to examine small for gestational age (SGA). Linear mixed models were applied to prenatal PFCAs and childhood weight and height z-scores.In girls, prenatal perfluorononanoic acid (PFNA), perfluorodecanoic acid (PFDeA), perfluoroundecanoic acid (PFUnDA), and perfluorododecanoic acid (PFDoDA) concentrations were inversely associated with birth weight ; prenatal PFDeA and PFUnDA were associated with elevated odds of SGA; and PFDeA, PFUnDA, and PFDoDA were associated with lower average childhood height z-score. In boys, prenatal PFNA, and PFDoDA were associated with reductions in height at certain ages in childhood, but not with size at birth.Prenatal exposure to long-chain PFCAs may interfere with fetal and childhood growth in girls, and childhood growth in boys.Wang Y, Adgent M, Su PH, Chen HY, Chen PC, Hsiung CA, Wang SL. 2016. Prenatal exposure to perfluorocarboxylic acids (PFCAs) and fetal and postnatal growth in the Taiwan Maternal and Infant Cohort Study. Environ Health Perspect 124:1794–1800; http://dx.doi.org/10.1289/ehp.1509998 They have been used widely in consumer products such as lubricants, polishes, water-repellant coatings for paper and textiles, fire-fighting foams, and cookware since the 1950s (Perfluorocarboxylic acids (PFCAs) are a class of synthetic organic chemicals, each consisting of a chain of fluorinated carbons, but varying in chain length , 2012b. he 1950s . PFCAs phe 1950s . Human ehe 1950s .Leading chemical manufacturers began efforts to reduce PFOA emissions and use around 2000 . In subsPFCAs are suspected developmental toxicants. Animal studies suggest PFCAs induce adverse effects on fetal/neonatal mortality and growth , particularly in response to high-dose exposures . In humaThe Taiwan Maternal and Infant Cohort Study is a longitudinal cohort study designed to examine prenatal environmental exposures and growth and development in infants and children . BetweenPrenatal exposure to PFOA and 4 long-chain PFCAs  was estimated from third-trimester maternal serum PFCA concentrations. Analytical methods have been described previously . In briez-scores by using sex and age-specific Taiwan national references (Birth weight (kg), length (cm), and head circumference (cm) were measured soon after delivery by nurses in the clinics. Small for gestational age (SGA) was defined as birth weight below the 10th percentile for the gestational age by sex using 1998–2002 Taiwan nationwide singleton birth weight charts . Weight ferences .n = 100,000) using these parameters. From this distribution, we took the mean of all values ≤ LOQ as our imputed value (E[x|x ≤ LOQ]). This method has been shown to attenuate the bias related to exposures under LOQ (r). Then we constructed linear regression models to examine each PFCA’s association with birth weight, length, and head circumference; and logistic regression models to assess associations with SGA. To fully assess the exposure, we also modeled prenatal PFCA concentrations as quartiles for continuous birth outcomes .We conducted all the analyses among full-term infants, separately by sex. The distributions of prenatal PFOA and long-chain PFCAs were right-skewed, so we used geometric means and percentiles to describe their distributions. For values below the LOQ, we imputed the expected values (E) conditional on being below the LOQ (E[x|x ≤ LOQ]). Specifically, for each PFCA, we estimated mean and variance parameters for a log normal distribution, based on observed percentiles of PFCA concentrations. We then simulated a log normal distribution , and annual family income . All analyses were conducted using a complete case approach. We did not adjust for smoking or alcohol consumption because few women smoked (2%) or consumed alcohol during pregnancy (1%). Because maternal weight was not routinely measured, our measurement of maternal weight gain during pregnancy (term weight – prepregnancy weight) was restricted to mothers for whom term weight measures were acquired (n = 109). Maternal weight gain and gestational age were additionally included in sensitivity analyses. Finally, we used Student t-tests to compare means and chi-square tests to compare frequencies between children who were included in our analysis (n = 223) and those who were not (n = 207). All the statistical analyses were performed in SAS , and all significance levels were set at α = 0.05.In all the models, the estimated coefficients (β) or odds ratios (OR) correspond to the unit change in the outcome per 1 ln-unit increase of PFCA concentration (ng/mL). To reduce bias due to potential confounders, we adjusted models for maternal age at delivery (years), education , previous live births , self-reported prepregnancy body mass index , 5 years , 8 years , and 11 years . n = 207), mothers who were included (n = 223) were more likely to have had a previous birth and, on average, gained less weight during pregnancy (see Table S1). Other characteristics, including prenatal serum PFCA concentrations, did not differ significantly between the groups. All PFCAs were detected in > 70% of the prenatal serum samples (r ≥ 0.56) but had only low to moderate correlation with PFOA (r ≤ 0.34) (see Table S2).In the present study, we included 223 full-term infants (106 girls and 117 boys) at birth, and conducted follow-up at ages 2 years , –0.15 , and –0.30 , respectively. No significant interactions between prenatal PFCAs and age at follow-up were found (p > 0.10); however, age-specific estimates suggest that reductions in girls’ weight z-score were limited to the time of birth and null at later ages , –0.14 , and –0.25 , respectively. Interaction terms for prenatal PFCA concentration-by-age were not significant (p > 0.10). For each PFCA, the magnitude of association was generally consistent across childhood . For height z-scores, however, we observed significant interactions with follow-up age for prenatal PFNA (p = 0.02) and PFDoDA exposure (p = 0.07). In general, age-specific estimates became increasingly negative with increasing age for all PFCAs  and delivery (n = 105)  were inversely, but not significantly, associated with birth weight, length, and head circumference; PFNA was associated with an increase in birth length and a decrease in ponderal index (n = 230) observed no association between preconception maternal PFOA, PFNA, or PFDeA and birth size  . In thisal index . Last, arth size .Despite strong geographic and temporal similarities in study samples, our study and the study by z-scores were imprecisely estimated due to loss to follow-up over time, and thus should be interpreted cautiously.In addition to birth outcomes, our study also observed an inverse association between long-chain PFCAs and average childhood height among girls, and height among boys at later ages of follow-up. We also observed a significant inverse association between prenatal long-chain PFCAs and girls’ average childhood weight, but this was attributed mainly to associations at birth. To our knowledge, no epidemiologic studies to date have assessed childhood growth in relation to prenatal exposure to long-chain PFCAs. However, a recent experimental animal study found that exposure to PFNA during gestation reduced pups’ postnatal weight in a dose-dependent manner . It is p9, C10, and C11 PFCAs are more hydrophobic and are less readily excreted through urine than C8 compounds  PDFClick here for additional data file.