Studies testing whether systemic inflammation might lie on the causal pathway between aberrant fetal and post-natal growth patterns and later cardiovascular disease have been inconclusive, possibly due to the use of single markers of unknown predictive value. We used repeated measures of a comprehensive set of inflammatory markers to investigate the relationship between early life measures and systemic inflammation in an African population.n = 320). Fasting levels of eight inflammatory markers  were measured, and potential confounding factors recorded. The association between early life measurements and systemic inflammation was assessed using regression analysis.Individuals born in three rural villages in The Gambia, and for whom early life measurements were recorded, were traced (Levels of most markers were unrelated to early growth patterns. In analyses adjusted for age and sex, more rapid growth between birth and 3 months of age was associated with higher levels of fibrinogen, orosomucoid, and sialic acid. These relationships persisted after further adjustment for body mass index but after full adjustment only the association with fibrinogen remained.This study provides little evidence that size at birth or growth in early infancy determine levels of inflammatory markers in young Gambian adults. Am. J. Hum. Biol. 25:457–464, 2013. © 2013 Wiley Periodicals, Inc. There is evidence that size at birth and rate of development during early life, in particular low birth weight   in West Kiang, The Gambia (West Africa) between 1976 and 1987 and for whom birth weight was recorded as part of the United Kingdom Medical Research Council (MRC) Keneba Antenatal Scheme. Details of this scheme and the research setting are available elsewhere  was recorded by the resident paediatrician or midwife to the nearest 10 g and within 72 h of birth. Gestational age was assessed using the score of Dubowitz et al. . Low birBlood samples were collected after an overnight fast and then centrifuged for 20 min at 3,000 rpm and 4°C within 1 h of collection and immediately frozen to −80°C. Serum samples were allowed to clot at room temperature, and then centrifuged and processed as described for plasma. Analyses of CRP, serum amyloid A, α 1-antichymotrypsin (ACT), orosomucoid, interleukin-6 (IL-6), sialic acid and neopterin, were performed at MRC Human Nutrition Research, Cambridge, UK. Fibrinogen was measured at Addenbrooke's Hospital Clinical Laboratory, Cambridge, UK.CRP was measured using a high-sensitivity particle-enhanced turbidimetric immunoassay  on a Dimension ARX Analyzer . The assay has a lower detection level of 1.1 mg/l. Plasma serum amyloid A was measured in duplicate using the enzyme-linked immunosorbent assay (ELISA) principle . Fibrinogen was measured using the Clauss assay. Plasma orosomucoid was measured using an immunoturbidimetric method  adapted for use on the Dimension ARX Analyzer . ACT was measured using an immunochemical assay  adapted for use on the Hitachi 912 analyzer . IL-6 was measured in duplicate using a high-sensitivity ELISA principle . Sialic acid was measured by a colorimetric enzyme assay  and adapted for use on the Hitachi 912 Analyzer . Serum neopterin was measured in duplicate using a competitive enzyme immunoassay principle . Interassay coefficients of variation were <9.6% for all analyses.2) was calculated and categorized using standard cut-offs  and height to the nearest mm using a portable stadiometer . BMI  and haemoglobin (g/dl) using a Medonic CA 530 Oden 16 Parameter System Haemoglobinometer .Participants were only enrolled if considered ‘healthy’ at the time of recruitment, based on a screening questionnaire collecting data on recent clinic visits, current medication use, appetite and recent weight loss. Axillary temperature was also recorded. A thick film was prepared from whole blood to look for the presence of malaria parasites. The remaining whole blood was used to measure white blood cell (wbc), lymphocyte, granulocyte and monocyte counts -cholesterol, triglyceride and leptin levels were measured at MRC Human Nutrition Research, Cambridge, UK. Plasma glucose concentration was measured using an adaptation of the hexokinase-glucose-6-phosphate dehydrogenase method . Impaired fasting glucose (IFG) was defined by a fasting plasma glucose ≥6.1 and ≤6.9 mmol/l and type 2 diabetes by a level ≥7.0 mmol/l . Smoking status was analyzed as ever or never smoked due to the small number of current smokers in the study population. Additional data were collected, by a female fieldworker, on whether women used oral or injectable hormonal contraceptives.Markers of systemic inflammation and infectious disease status were assessed at two time points [baseline (Day 0) and 2-weeks (Day 14)]. In a random sub-sample of 15 women these levels were further assessed at Day 28 to investigate the reliability of a single measurement to characterise habitual systemic inflammatory status. All other measurements were collected at baseline only. Baseline data were collected between 23 February and 1 June, 2-week data between 9 March and 15 June and 4-week data between 24 March and 30 March, 2006.e transformed to normality. A loge transformation of serum amyloid A failed to produce a normal distribution and it was necessary to add 100 to each serum amyloid A value and then take the loge transformation to obtain a normal distribution.A total of 209 (65.3%) participants had a CRP and 266 (83.1%) an IL-6 measurement below the minimum assay detection level . CRP and IL-6 were therefore analyzed as binary variables . Orosomucoid, ACT, sialic acid and neopterin were loge) for those variables which were transformed to produce a normal distribution. All SD scores were sex-specific. SD-scores could not, however, be generated for anthropometric, body composition and leptin measurements as a number of these measures could not be transformed to produce the normal distribution necessary to generate SD scores. The effect of anthropometric, body composition and leptin measurements on levels of inflammatory markers were examined in males and females separately because of the strong sex-differences in these measures observed within the study population.The effects of potential confounding factors  are presented as absolute changes.The associations between birth weight, weight at 1 year and postnatal growth with levels of inflammatory markers were analyzed using logistic regression for CRP and IL-6 and linear regression for all remaining inflammatory markers according to the analytical approach recommended by Lucas et al. . This apn = 392) of the 452 individuals traced of whom 72 (18.4%) declined to participate. The majority of those who declined were male . The remaining 320 individuals represent 70.8% of those traced and available for study (n = 452).P = 0.0001) and a slightly higher percentage were male . Mean birth weight, gestational age, change in SD score from birth to 3 months and weight at 1 year and percentage born during the hungry season or low birth weight were not different between participants and nonparticipants.Compared with nonparticipants, study participants were younger  age 22.2  vs 23.0  years; P = 0.02). Gestational age ranged from 32.0 to 41.6 weeks with 30 participants (11.2%) born premature (<37 weeks gestation). The majority of participants had a BMI within the normal range although the percentage overweight, or centrally obese, was considerably higher in females compared to males (P < 0.001 for both). As expected in young Gambian adults there was a low prevalence of hypertension (<3%), IFG (<1%) and type 2 diabetes (<1%). Less than 1% of the study population had asymptomatic malaria. There was a clear sex-difference in tobacco use with no females, compared with 30% of males reporting ever using tobacco regularly. Few women (3.3%) used hormonal contraceptives.P = 0.004), orosomucoid levels increased by 3%  and sialic acid levels by 2% (95% CI 1–4%); P = 0.003).In regression analyses, each adjusted for age and sex, there was no evidence that birth weight, low birth weight, weight at 1 year or season of birth predicted levels of systemic inflammatory markers. There was evidence that a higher change in SD score from birth to 3 months was associated with higher levels of fibrinogen, orosomucoid and sialic acid but no association was observed with the remaining five markers. For each one unit increase in change in SD score fibrinogen levels increased by 0.11 g/l  and showed a weak positive association with sialic acid (1.00 (95%-0.05 to 2.92)%; P = 0.06) but any association with orosomucoid levels was removed. For each inflammatory marker, multiple regression analyses were used to adjust for measures of location, chronic disease, infectious disease and lifestyle . The only exception to this was that adjustment for DXA, total or percent fat, rather than BMI, forced the association between postnatal growth and levels of sialic acid toward the null.The current study aimed to test the hypothesis that systemic inflammation is young adulthood is predicted by early life exposures. Using a comprehensive set of markers in a cohort of 320 young Gambian adults, we have found little evidence for an effect of a number of early life parameters on later inflammatory status. This is the first published study to investigate this association in an indigenous African population. The inclusion of a range of early life measures, in particular early postnatal growth, and a large and varied number of markers to assess systemic inflammation considerably widens the evidence base in this field.To date, the only published data supporting an association between prenatal exposures and systemic inflammation comes from studies looking at the relationship between birth weight and either CRP or fibrinogen. Previous studies in infants and young children have reported no association between birth weight and levels of CRP  birth cohort suggest that rapid weight gain across the life course predicts higher CRP at age 23 years  [the rains from June to October coincide with an increase in disease transmission . Comparisons between participants and nonparticipants however suggest no differences between the two groups in the exposure variables available for analysis. In addition, it is important to highlight that this study was investigating comparisons within-subjects and it is unlikely that in nonparticipants any associations between early life factors and levels of inflammatory markers would be in the opposite direction to that observed in participants.A critical weakness of the study is that the population measured represents only 41% of those who met the initial study criteria (It has been suggested that the association between birth weight and other determinants of cardiovascular disease (e.g. blood pressure) amplifies with age; the relatively young age of the study population (19–30 years) may have prevented any association between early life events and levels of inflammatory markers being observed. Furthermore, the study participants were relatively lean and therefore underlying associations may exist but remain hidden.A technical limitation of the study design was that despite using high sensitivity kits the CRP and IL-6 assays were not sufficiently sensitive in the lower ranges. In total, 65 and 83% of participants had levels of CRP and IL-6 below the assay ranges, respectively; CRP and IL-6 were subsequently analyzed as binary variables which limited the power to observe any association with early life factors. No evidence-base was available with which to decide the cut-offs for generating the binary variables. It is possible that the cut-offs used in this study (above and below the lowest assay cut-off) also limited observations on the effect of early life factors on levels of CRP and IL-6.The main hypothesis of this study was that early life programming of systemic inflammation was a mechanism to explain the association between poor early life growth and increased risk of adult CVD; this study provided little evidence to support this hypothesis among young, lean Gambians.


We assessed anthropometric status, breastfeeding duration, morbidity, and mortality outcomes during the first four years of life according to gestational age, in three population-based birth cohorts in the city of Pelotas, Southern Brazil.Total breastfeeding duration, neonatal mortality, infant morbidity and mortality, and anthropometric measures taken at 12 and 48 months were evaluated in children of different gestational ages born in 1982, 1993 and 2004 in Southern Brazil.Babies born <34 weeks of gestation and those born between 34–36 weeks presented increased morbidity and mortality, were breastfed for shorter periods, and were more likely to be undernourished at 12 months of life, in comparison with the 39–41 weeks group. Children born with 37 weeks were more than twice as likely to die in the first year of life, and were also at increased risk of hospitalization and underweight at 12 months of life. Post-term infants presented an increased risk of neonatal mortality.The increased risks of morbidity and mortality among preterm (<37 weeks of gestation) and post-term (>41 weeks) are well known. In our population babies born at 37 also present increased risk. As the proportion of preterm and early term babies has increased markedly in recent years, this is a cause for great concern. During pregnancy, every week counts for growth and development, and the decision to interrupt gestation should be based on weighing the risks that the baby is facing in utero against the risks of preterm delivery.The current definition of preterm births – less than 37 completed weeks of gestation, or three weeks before the date of delivery – was adopted by the World Health Organization  in 1975.However, a number of recent studies from the United States and Europe show that babies born with 37 and 38 weeks of gestation also run increased risk, in comparison with those born with 39–41 weeks .Because we were unable to find any studies from low or middle income countries on this topic, we decided to study the distribution by gestations age of several indicators of mortality, morbidity, anthropometric status and breastfeeding duration, with special emphasis on children born with 37 and 38 weeks of gestation. These outcomes were measured during the first four years of life in three population-based birth cohorts in the city of Pelotas in Southern Brazil.During 1982, 1993 and 2004, birth cohorts representing all birth to mothers residing in the urban area of Pelotas, Southern Brazil, were enrolled. The three studies relied on primary data collection with the same methodology,-18 incluData from the perinatal study and from follow-up home visits at 12 and 48 months were used in the present study. In the 1982 and 2004 cohorts, we attempted to locate all babies recruited in the perinatal phase. In the 1993 cohort we sought all low birth weight babies and a 20% sample of the remained; weighted data analyses were used to reproduce the total cohort [Follow-up losses at 12 and 48 months visits were equal to 20.7% and 15.9% for the 1982 cohort, 6.6% and 12.8% for the 1993 cohort, and 5.9% and 8.2% for the 2004 cohort, respectively.During each home visit mothers were interviewed by trained personnel who also conducted anthropometric measurements of the children. Tanita electronic scales  with 100g precision were used for weight, the scales being calibrated on a weekly basis. Lengths and heights were measured with portable infantometers with 1mm precision. Birth length was not available for the 1982 birth cohort.Gestational age estimation was based on the date of the last menstrual period (LMP). Cases of unknown or implausible LMP were treated as missing information. They corresponded to 15.6% of the births – 21% in 1982, 11% in 1933, and 7.3% in 2004. Gestational age was categorized as <34, 34–36, 37, 38, 39–41 and ≥42 completed weeks.Mortality surveillance was carried out actively through regular visits to all hospital, cemeteries, state vital registration services and the city's health department. Infant and neonatal mortality were defined as the deaths of live-born infants in the first 364 or28 days of live, respectively. Deaths in these periods were expressed per thousand live births: the infant (IMR) and neonatal mortality rates (NMR).Information regarding breastfeeding was reported by the mother. Total breastfeeding duration (in months and days) was collected at each follow-up. The earliest available information on breastfeeding cessation was used to reduce recall bias.Information on hospital admissions in the first 12 months of life was obtained from the mothers and was restricted to re-admissions after the newborn had been discharged from the maternity hospital.The following anthropometric indices were calculated: weight-for-age (W/A), height-for-age (H/A) and weight-for-height (W/H) z-scores at 12 and 48 months, using the World Health Organization (WHO) growth standards . UnderweThe following factors, measured in the perinatal period, were considered to be potential confounders of the association between gestational age and each outcome. Family income in the month prior to delivery was analyzed in quintiles; maternal schooling , maternal age (completed years) and height (in centimetres) were analyzed as continuous variables. Women who were single, widowed, divorced or lived without a partner were classified as single mothers. Parity was defined as the number of previous pregnancies resulting in a live birth or a late fetal death. Smoking during pregnancy, regardless of the number of cigarettes, was categorized as yes or no. Breastfeeding was not considered as a confounder, because it could not influence gestational age; in fact, it constitutes a potential mediator in the association between gestational age and the other outcomes.We tested heterogeneity between the three cohorts in terms of the association between gestational age and each outcome. Because no significant interactions were found (p<0.10), we pooled the data from the three cohorts.t-test or x2 statistics to study crude associations between gestational age and each outcome. Confounder-adjusted analyses included logistic regression analysis for categorical outcomes and multiple linear regression for continuous outcomes. Analyses were carried out using Stata software, v. 11.0.We used The study protocols of the three cohort studies were approved by the Medical Ethics Committee of the Federal University of Pelotas, which is affiliated with the Brazilian Federal Medical Council. In the 1982 and 1993 cohort studies, verbal consent to participate in the study was obtained from mothers. In the 2004 cohort study, mothers were provided written informed consent.We examined separately the outcomes for each birth cohort, and the tables can be found as additional material files , but also amongst those born at 37 weeks is a cause for concern. In Pelotas, where the whole gestational age curve has shifted to the left in the last three decades, and the prevalence of preterm births increased from 6.2% in 1982 to more than 15% in 2004 . Births Finally, it is not possible to ascertain in an observational study if the increased risks observed among infants born before full maturity were due to early exposure to the extra-uterine environment, or damages produced by maternal conditions that may also have produced the untimely birth, or even a combination of both situations. A previous study suggests that preterm birth and exposure to maternal medical conditions are independent risk factors for neonatal morbidity, with the former playing a stronger role .This study supports the literature from high-income countries -15 in stH/A: Height-for-age z-score; IMR: Infant mortality rate; NMR: Neonatal mortality rate; W/A: Weight-for-age z-score; W/H: Weight-for-height z-scores; WHO: World Health Organization.The authors have indicated they have neither financial relationships relevant to this article to disclose nor any conflict of interests.FCB, JLDR, and CGV contributed to the concept and the design of the study, and writing of the manuscript. AM, SCD, AJDB, ISS, and DM contributed to the analysis of the data. All authors contributed to the interpretation of the data and revising the manuscript, and final approval for publication. FCB and CGV are the guarantors. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2431/12/169/prepubTable 1. Frequency of different outcomes during the first four years of life according to gestational age. Pelotas (Brazil) 1982 Birth Cohort. Table 2. Adjusted* relative risks  and beta coefficients  of different outcomes according to gestational age . Pelotas (Brazil) 1982 Birth Cohort.Click here for fileTable 1. Frequency of different outcomes during the first four years of life according to gestational age. Pelotas (Brazil) 1993 Birth Cohort. Table 2. Adjusted* relative risks  and beta coefficients  of different outcomes according to gestational age (reference group= 39-41 weeks). Pelotas (Brazil) 1993 Birth Cohort.Click here for fileTable 1. Frequency of different outcomes during the first four years of life according to gestational age. Pelotas (Brazil) 2004 Birth Cohort. Table 2. Adjusted* relative risks  and beta coefficients  of different outcomes according to gestational age (reference group= 39-41 weeks). Pelotas (Brazil) 2004 Birth Cohort.Click here for fileTable 1. Number of cases analyzed for each outcome and p-value for the regression analyses for interactions between each outcome by cohort and gestational age groups. Pelotas (Brazil) Birth Cohorts 1982, 1993, 2004.Click here for file


Related research article Olive V, Sabio E, Bennett MJ, DeJong CS, Biton A, McGann JC, Greaney SK, Sodir NM, Zhou AY, Balakrishnan A, Foth M, Luftig MA, Goga A, Speed TP, Xuan Z, Evan GI, Wan Y, Minella AC, He L. 2013. A component of the miR-17-92 polycistronic oncomir promotes oncogene-dependent apoptosis. eLife2:e00822. doi: 10.7554/eLife.00822Image B-cell lymphomas overexpressing the entire miR-17-92 cluster (left) exhibited more apoptotic cells than those overexpressing the miR-17-92 cluster without miR-92a (right).bona fide oncogenes and tumour suppressors. In particular, the miR-17-92 cluster—a group of six microRNAs all encoded by the same gene—has been studied extensively since researchers first discovered that it is overexpressed in Burkitt’s lymphoma, diffuse B-cell lymphoma and multiple types of solid tumours  as the key oncogenic component of the cluster in this model  exhibited significantly stronger oncogenic activity than the full cluster in Eμ-myc B cells. Similarly, mutations that inactivated miR-92a strongly enhanced the oncogenic effects of the full cluster. In pre-malignant B cells from Eμ-myc mice, and also in mouse embryonic fibroblasts, the expression of miR-92a triggered programmed cell death (apoptosis), thereby counterbalancing the effects of miR-19 (which suppresses apoptosis). Intriguingly, however, the overexpression of miR-92a also increased the proliferation of fibroblasts and B cells. This dichotomy can be at least partially explained by the ability of miR-92a to reduce expression of the gene Fbw7, which encodes an enzyme that facilitates the degradation of the MYC protein. Thus, the increase in MYC levels resulting from miR-92a expression simultaneously stimulates proliferation and engages the p53 tumour suppressor pathway, resulting in apoptosis (Olive, Sabio, He and co-workers have now further delineated the contributions of individual microRNAs within poptosis , left.miR-92a-Fbw7-MYC axis clearly contributes to the apoptosis induced by miR-92a, it is unlikely that this mechanism can fully explain the ability of this microRNA to induce cell death. For example, elevation of MYC protein in fibroblasts does not cause apoptosis as efficiently as expressing miR-92a. Additionally, Olive et al. did not demonstrate that Eμ-myc lymphoma cells with enforced miR-92a expression have higher MYC levels than lymphoma cells without enforced miR-92a. Further studies will be necessary to establish precisely how this pathway influences the growth and development of B-cell lymphomas, and whether miR-92a leads to apoptosis via other pathways.While the miR-19 to miR-92a determines the overall oncogenic potency of the miR-17-92 cluster. Interestingly, they found that miR-19 is more highly induced than miR-92a in pre-malignant and malignant Eμ-myc B cells compared to normal B cells. A similar phenomenon is evident in human Burkitt’s lymphoma cell lines. These observations suggest that the processing of miR-19 and miR-92a from the miR-17-92 primary transcript or the stability of the individual mature miRNAs is differentially regulated in normal B cells compared to lymphoma cells. Elucidation of this mysterious post-transcriptional regulatory pathway, which itself may be oncogenic, will undoubtedly be a priority for future work.Olive et al. suggest that the ratio of miR-92a depends on the presence of an intact p53 pathway, it is important to consider the effects of expressing this microRNA when p53 is absent, as is the case in a large fraction of human tumours. In the absence of p53, the ability of miR-92a to enhance levels of the protein MYC will undoubtedly serve a pro-tumorigenic agenda. Indeed, miR-92a has been shown to promote the proliferation and survival of liver and colon cancer cell lines (miR-17-92 (miR-92a may result in a feed-forward loop that leads to even greater activation of MYC. In such a setting, miR-92a may work with, rather than against, miR-19, with both of these microRNAs promoting the formation and growth of tumours through independent pathways (Given that the pro-apoptotic activity of ll lines . MoreoveiR-17-92 , it is cpathways , right.miR-199-214 cluster work together to promote cardiac remodelling (miR-17-92 cluster or if it can be generalized to other microRNAs and other species.This work also begins to address a broader question: why have microRNA clusters evolved? One might expect that co-transcribed microRNAs would generally work in concert and, indeed, there are many such examples. For instance, both microRNAs within the odelling . Neverth


Globally, 11% of infants are born preterm. In adulthood, individuals born preterm are at increased risk of cardiovascular morbidity and mortality, but the mechanistic basis of this remains unknown. Clinically overt cardiovascular disease may be preceded by altered cardiac autonomic activity characterised by increased sympathetic activity and/or reduced parasympathetic activity. Thus, altered cardiac autonomic activity in survivors of preterm birth may underlie later cardiovascular risk.To investigate the impact of gestational age on cardiac autonomic activity in juvenile and adult sheep.Singleton-bearing ewes were randomised antenatally to spontaneous term birth  or corticosteroid induced preterm birth . Cardiac autonomic modulation was assessed using heart rate variability analysis in juvenile and adult offspring. Preterm birth in adult males was associated with altered sympatho-vagal modulation  and reduced parasympathetic modulation . Within the range of term birth, each one-day increment in gestational age was associated with a decrement in LFnu in juvenile females and with a decrement in LFnu and LF/HF ratio, but an increment in RMSSD and NN50 in adult females. Cardiac autonomic function in adult sheep is affected in a sex-specific manner by gestational age at birth, even within the term range. Altered cardiac autonomic function may contribute to increased later cardiovascular morbidity in those born preterm. As preterm birth rates and survival following preterm birth increase, it is becoming increasingly apparent that preterm birth is accompanied by later cardiovascular morbidity: increased blood pressure ,2, alterThe fetal autonomic nervous system increases in complexity as gestation advances, with reducing sympathetic and increasing parasympathetic activity . In the We hypothesized that gestation length at birth would alter autonomic balance in juvenile and adult sheep and that this might provide a mechanistic basis for increased cardiovascular morbidity in those born preterm. Ethics statement: Ethical approval for this study was obtained prospectively from the University of Auckland Animal Ethics Committee (AEC 628). Time-mated, singleton-bearing Romney ewes (n=133) were generated . At day PT ewes received 0.25mg/kg dexamethasone sodium phosphate  by intramuscular injection on days 135 and 136 with birth of the lamb on d137. TC ewes received no interventions during pregnancy. ® system  from standing, conscious sheep in a quiet, temperature and light-controlled  environment to which they had been extensively habituated. To control for any effect of diurnal variation in autonomic tone [At 5 months’ (juvenile) and 16 months’ (adult) corrected postnatal age, electrocardiograms (ECGs) were obtained using a PowerLabad libitum. Subcutaneous ECG needles  were placed on the sternum, over the left scapula and the left flank and secured in place with adhesive [Animals were studied while housed in metabolic crates to which they were also well habituated, with standard sheep pellets and water available adhesive . After aadhesive . These fadhesive , reflectadhesive . ECGs weAs cardiovascular disease risk is known to be different between the sexes, males and females were analysed separately using JMP v. 8 . Non-parametric data were log transformed prior to analysis. Data were analysed by factorial ANOVA with Tukey post hoc correction. Associations between continuous variables were examined by regression analysis within each birth group, in same sex animals. Multivariate regression models were adjusted for the effect of: birthweight z-score (birthweight - mean birthweight for gestation and sex / SD of mean birthweight for gestation and sex), age and weight (juveniles) or weight-for-length (weight (kg) / crown-rump length (m)) in adults as a better measure of soft tissue mass relative to skeletal growth. Measures of length were not available for juvenile sheep. At birth, but not in juvenile or adult life, same-sex PT lambs were smaller and lighter than TC lambs . In males, PT sheep had altered indices of sympatho-vagal modulation as juveniles and adults, with reduced indices of parasympathetic modulation in adulthood, but no difference in heart rate at either age . In femaTC sheep had a gestational age range of 142-156 days . In females, each one-day increment in gestational age was associated in juveniles with a decrement in LFnu and in adulthood with a decrement in LFnu and LF/HF ratio, but an increment in RMSSD and NN50 . Our data demonstrate that both preterm birth and gestation length within the term range result in sex-specific effects on autonomic balance in juvenile and adult sheep, with reduced gestation length associated with altered sympatho-vagal balance. Although HRV does not represent direct measurements of autonomic nerve activity, it is a sensitive technique for estimation of overall autonomic balance and is predictive of cardiovascular morbidity. Tonic elevation of sympathetic activity and / or parasympathetic withdrawal may precede development of clinically overt hypertension ,21 and cGestational age and preterm corticosteroid exposure both may be implicated in early postnatal autonomic function and separation of the two factors is challenging. Steroid-naïve preterm birth in experimental animals is associated with significant mortality  and, in per se is unlikely to be the sole mediator of the differences in HRV between preterm- and term-born sheep. Preterm birth in lambs is not accompanied by the increase in HR, BP and renal sympathetic nerve activity observed in term-born lambs unless the ewe receives antenatal corticosteroid treatment . Howeverin situ to control for the effect of variation in oestrous cycling between animals. The impact of gestational age at birth on age-related changes in sex hormones was not assessed in this study, but may be one mechanism through which gestational age modifies later sympatho-vagal balance. Further studies are needed to define the interaction between normal cyclical changes in sex hormones in animals born preterm on modulation of autonomic tone. It is also possible that in our female sheep the differences in baseline heart rate masked differences in HRV [The emergence of sexually dimorphic HRV characteristics is congruent with the sex-specific differences in manifestation of cardiovascular disease . Sexualls in HRV . HoweverOur data describe the effect of gestational age and preterm birth in singleton animals only. In humans, although preterm birth is common in twins and higher order multiples, singletons account for approximately 75% of the preterm population . DiffereFinally, the large number of animals studied is a strength of this study but also limited us to the use of a tethered ECG system with relatively short ECG recordings compared to those that can be acquired through a telemetric system. Future studies will include ECG recordings throughout a 24 hour period and also blood pressure recordings to better define the low frequency modulation of HRV and better characterise the contribution of the sympathetic nervous system to overall autonomic balance. Despite these limitations, we believe that these data represent a novel insight into a late autonomic dysfunction arising from preterm birth. Preterm birth in sheep results in altered sympatho-vagal modulation of autonomic activity in adulthood in a sex-specific manner. This may contribute to the emergence of adult cardiovascular morbidity in humans born preterm.


Preterm birth is a major contributor to neonatal morbidity and mortality and its rate has been increasing over the past two decades. Antidepressant medication use during pregnancy has also been rising, with rates up to 7.5% in the US. The objective was to systematically review the literature to determine the strength of the available evidence relating to a possible association between antidepressant use during pregnancy and preterm birth.a priori design was to group studies by level of confounding adjustment and by timing of antidepressant use during pregnancy; we used random-effects models to calculate summary measures of effect.We conducted a computerized search in PUBMED, MEDLINE and PsycINFO through September 2012, supplemented with a manual search of reference lists, to identify original published research on preterm birth rates in women taking antidepressants during pregnancy. Data were independently extracted by two reviewers, and absolute and relative risks abstracted or calculated. Our rd trimester use. Controlling for a diagnosis of depression did not eliminate the effect. There was no increased risk [1.16 (0.92–1.45)] in studies that identified patients based on 1st trimester exposure. Sensitivity analyses demonstrated unmeasured confounding would have to be strong to account for the observed association.Forty-one studies met inclusion criteria. Pooled adjusted odds ratios (95% CI) were 1.53 (1.40–1.66) for antidepressant use at any time and 1.96 (1.62–2.38) for 3nd and 3rd trimesters, although the possibility of residual confounding cannot be completely ruled out.Published evidence is consistent with an increased risk of preterm birth in women taking antidepressants during the 2 Preterm birth is a major clinical problem throughout the world. It is the leading cause of infant mortality: approximately 75% of all perinatal deaths occur among preterm infants In many developed countries, the use of antidepressant medications has increased sharply between 1996 and 2005, and now surpasses antihypertensives as the most commonly prescribed drug class in ambulatory care Numerous studies, of varying size and quality, have examined the effects of antidepressant medication use on pregnancy outcomes, including preterm birth. They differ in terms of the timing of the antidepressant exposure during pregnancy and adjustment for potential confounding variables, including lifestyle factors, co-morbidities, and the severity of the underlying psychiatric illness. The extent to which such differences contribute to variability in findings remains to be elucidated. The objective of this review was to determine the strength of the available evidence relating to a possible association between antidepressant use during pregnancy and preterm birth, and to assess this relationship in terms of (1) the timing of the antidepressant use studied, and (2) attempts to control for the possible confounding effects of depression itself.To identify all available studies on the topic of antidepressant medication use during pregnancy and preterm delivery, we performed a computerized search in PUBMED, MEDLINE, and PsycINFO using the key words: (“antidepressant*” or “tricyclic antidepressant*” or “selective serotonin reuptake inhibitor*” or “serotonin-norepinephrine reuptake inhibitor*”) and (“preterm birth*” or “preterm deliver*” or “pregnanc*” or “pregnancy complication*”). The databases were searched from their inception through September 12, 2012. Reference lists of selected articles were also searched to identify additional studies that reported on preterm births and antidepressant exposure.Studies were included if they identified a group of pregnant women exposed to antidepressants at some point during their pregnancy as well as a comparison group, and reported on preterm birth rates, irrespective of whether preterm birth was a pre-specified study endpoint or one of several pregnancy characteristics reported. No restrictions were imposed on study size or design.Once relevant studies were identified, for each study two investigators with clinical and epidemiologic expertise  independently abstracted preterm birth rates in the group(s) exposed to antidepressants and in the comparator group(s). When these were not expressed as percentages in the manuscript, they were calculated. Relative risks  were also taken directly from the manuscript. Whenever available, preference was given to relative risks adjusted for potential confounding variables. When relative risks were not reported in the manuscript, we estimated the unadjusted odds ratio and its corresponding 95% confidence limits  based on the available information. When relative risks were only presented graphically, we contacted the authors to obtain the corresponding numerical estimates. Types of antidepressants used, numbers of users and other pertinent exposure information , and potential confounders accounted for were also retrieved. Any discrepancies in data abstraction were resolved by consensus between the reviewers.2 tests a priori to group studies by timing of antidepressant exposure and level of confounding adjustment. We chose this approach to address the inherent issues of clinical and methodological diversity A funnel plot was examined for evidence of publication bias After an initial screen of the 1,477 studies identified through database searches, we identified 52 studies that met our predetermined criteria as possibly assessing the association between antidepressant use during pregnancy and preterm birth Forty-one studies All but one study defined preterm birth as an infant born before 37 weeks’ gestation, in accord with the WHO definition. Maschi et al considered infants born before 36 weeks of gestation as premature 2∶46 to 85%), but not across studies that provided unadjusted estimates (I2<25%). We conducted a random-effects meta-analysis of the adjusted estimates given the consistency in the direction of the effects Early’ includes studies in which women were known to have taken antidepressants early in pregnancy, typically in the first trimester. Some of these women continued antidepressant medication use during pregnancy whereas others discontinued. Studies classified under ‘Any time’ are those in which women were considered exposed irrespective of the specific time during pregnancy that medication was used, and studies where the timing was not specified. If multiple exposures were analyzed in a given study  for early exposure, and 1.44 (1.34–1.56) for exposure any time during pregnancy , Table 2The adjusted odds ratios for the 22 studies which accounted for potential confounding variables are shown in Early’ and ‘Any Time” categories defining the timing of antidepressant use, Late’ category which comprises studies in which women were known to use antidepressant medications late in pregnancy, generally in the third trimester. Results suggest an increased risk of preterm birth for all exposure-outcome combinations, with four exceptions. Calderon-Margalit and colleagues In addition to the previously defined ‘In general, associations appeared stronger for antidepressant use known to have occurred late in pregnancy  , Table 2A major concern about the validity of studies assessing the effect of antidepressant medications on preterm birth is the potential for confounding by indication. It has been hypothesized that the underlying depression and its severity, or the behaviors potentially associated with depression , rather than antidepressant medication, may themselves increase the risk of preterm birth A few studies indirectly addressed the issue of confounding by depression by comparing preterm birth rates in women who continued vs. those who discontinued their antidepressant medication during pregnancy. Comparing women who continued antidepressant use through the third trimester with those who discontinued, Chambers and colleagues estimated an OR of 4.8 for fluoxetine EC) and the outcome (RRCD) that would be required to fully explain the observed increased rate of preterm birth associated with antidepressant medication use during pregnancy (depression-adjusted OR = 1.61) if in truth no such increase existed. For an unmeasured confounder present in 25% of the population, relative risks ≥4 linking the hypothetical confounder to both antidepressant medication use and preterm birth would need to be present to fully explain the observed association, assuming 8% of pregnant women are exposed to antidepressants Visual inspection of the funnel plot reveals some asymmetry, suggesting smaller studies with negative associations might be under-represented in the literature . It shouThis systematic review of the literature concerning the relationship between antidepressant use and preterm birth identified 41 observational studies. Findings suggest the risk of preterm birth is increased in women treated with antidepressant medications during pregnancy, with pooled odds ratio estimates ranging between 1.16 (95% CI 0.92–1.45) and 1.96 (95% CI 1.62–2.38). The associations were stronger for antidepressant use later in pregnancy. Adjusting for a diagnosis of depression in most cases did not eliminate the effect, although the strength of the observed associations was somewhat attenuated. Sensitivity analyses demonstrated that very strong risk factors of preterm birth that are fairly imbalanced among exposure groups and independent of the adjusted confounders must be unmeasured and uncontrolled to explain the observed associations. Although it would be unlikely to miss such a strong single confounder, it is conceivable that several weaker confounders could have acted together to account for the apparent effect. Our findings are consistent with those from an earlier study which evaluated the association between prenatal antidepressant exposure and a range of adverse pregnancy and delivery outcome Preterm birth is a major problem worldwide First, antidepressant exposure in many pregnancies is not a “yes or no” phenomenon. Many women stop antidepressants when they discover they are pregnant, resulting in first trimester exposure only. Others stop, but may restart the medications later in the pregnancy. Still other patients do not take the medications for much of the pregnancy but start in the third trimester. The issue of classifying exposure is therefore complex. Several studies, for example, have been performed by Teratogen Information Services (TIS). Women who called the TIS and reported that they were taking antidepressants were classified as “exposed” in those studies. The advantages of this design include the fact that exposure is determined prior to outcome, eliminating recall bias, and a control group is readily identifiable–women who contact the TIS with exposure to another medication. Yet, there are also major drawbacks to this approach. Most women who contact TIS are concerned about the use of antidepressants during pregnancy; therefore many of them may not continue with the medication throughout the pregnancy. If exposure to antidepressants in the second and third trimesters is more likely to be associated with preterm birth (and available evidence suggests that this might be the case) then TIS studies that classify women as exposed solely on the basis of first trimester exposure would likely underestimate the association between antidepressant use and preterm delivery. Studies that rely on electronic healthcare databases, on the other hand, contain detailed information on filled prescriptions for antidepressant medications during the entire pregnancy. Automated pharmacy dispensing information is usually seen as the gold standard of drug exposure compared to self-reported information A major concern is the potential for confounding by the underlying depression and its severity, and associated poor health behaviors . More severely depressed women may be more likely to take antidepressants during pregnancy, and it has been suggested that it may be the depression itself that is causing the preterm birth and not the medication. Several of the studies in this systematic review made efforts to control for maternal depression and these studies continued to show increased rates of preterm birth in the antidepressant exposed pregnancies. The majority of studies we reviewed did not find increased preterm birth rates in depressed women unexposed to medication. However, despite these studies’ attempts to control for depression, it is likely that women with a diagnosis of depression who opt to continue treatment during pregnancy are inherently different from women with a depression diagnosis who discontinue treatment during pregnancy. It is therefore questionable whether these studies completely addressed confounding by indication severity. Nevertheless, the available data on whether depression itself is associated with preterm birth is inconsistent Some limitations of this review, resulting from limitations in the source data, should be noted. The studies included were heterogeneous in terms of design, size, exposure assessment, timing and nature of the exposure, and confounding adjustment. For example, in some studies, women classified as exposed were only those taking medication throughout the pregnancy, and controls were those who stopped antidepressant use before the second trimester, while in others the “exposed” were patients taking an antidepressant in the first trimester, who may have stopped by the second trimester. We tried to address this by presenting the results separately by level of adjustment and timing of the exposure, but assignment of studies to these categories is somewhat subjective, and heterogeneity within categories remains. Although we would have liked to simultaneously investigate the effects of some of these factors through meta-regression, this was not feasible due to an insufficient number of studies. Given that all but one study defined preterm birth as a dichotomous outcome , we were not able to examine the strength of the association between antidepressant medication use and very-, moderate- and mild-preterm birth respectively. Nevertheless, even late preterm birth is a significant contributor to poor neonatal outcomes The most rigorous method for determining an association between antidepressant medication and preterm birth would be a randomized controlled trial, and some have argued for this In conclusion, the findings from our review of the literature are consistent with an association between antidepressant use during pregnancy and preterm birth, although the possibility of residual confounding by depression severity cannot be completely ruled out based on the available evidence. Counseling of pregnant women must take into consideration the clinical circumstances of a given patient, the strength of the available evidence on the risks and benefits , and alternatives to medication use during pregnancy. While our study findings cannot prove causality, they reinforce the notion that antidepressants should not be used by pregnant women in the absence of a clear need that cannot be met through alternative approaches.Checklist S1Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) checklist.(DOC)Click here for additional data file.


Significant and persistent racial and income disparities in birth outcomes exist in the US. The analyses in this manuscript examine whether adverse birth outcome time trends and associations between area-level variables and adverse birth outcomes differ by urban–rural status.Alabama births records were merged with ZIP code-level census measures of race, poverty, and rurality. B-splines were used to determine long-term preterm birth (PTB) and low birth weight (LBW) trends by rurality. Logistic regression models were used to examine differences in the relationships between ZIP code-level percent poverty or percent African-American with either PTB or LBW. Interactions with rurality were examined.Population dense areas had higher adverse birth outcome rates compared to other regions. For LBW, the disparity between population dense and other regions increased during the 1991–2005 time period, and the magnitude of the disparity was maintained through 2010. Overall PTB and LBW rates have decreased since 2006, except within isolated rural regions. The addition of individual-level socioeconomic or race risk factors greatly attenuated these geographical disparities, but isolated rural regions maintained increased odds of adverse birth outcomes. ZIP code-level percent poverty and percent African American both had significant relationships with adverse birth outcomes. Poverty associations remained significant in the most population-dense regions when models were adjusted for individual-level risk factors.Population dense urban areas have heightened rates of adverse birth outcomes. High-poverty African American areas have higher odds of adverse birth outcomes in urban versus rural regions. These results suggest there are urban-specific social or environmental factors increasing risk for adverse birth outcomes in underserved communities. On the other hand, trends in PTBs and LBWs suggest interventions that have decreased adverse birth outcomes elsewhere may not be reaching isolated rural areas. Significant racial and income disparities in birth outcomes exist in the US  and haveHealth disparities across the urban–rural gradient also exist; both inner city and rural residents, compared to suburban, have poorer health outcomes . The trePrevious research has not determined the extent that area-level socioeconomic factors contribute to birth outcome disparities or whether rurality modifies this effect. Several studies have found significant area-level socioeconomic adverse birth outcome disparities remained after individual variables were included in the model ,17-19. HUnlike the rest of the US, both urban and rural areas in the Deep South have significant non-Hispanic African American populations, allowing for characterization of racial health disparities across the urban–rural gradient. Alabama is consistently among the states with the highest PTB and LBW rates in the US. In 2010, the PTB rate in Alabama was 15.6% compared to the national average of 12.0%, and Alabama’s African American communities have particularly high PTB and LBW rates .Here we examine whether contributions of individual or area-level factors to health disparities in the Deep South vary by urban–rural status over the last 20 years. We have two objectives: (1) determine whether 20-year temporal trends in PTB and LBW in Alabama vary by rurality and (2) determine if rurality modifies the effects of area-based racial composition or poverty with regards to individual-level risk for PTB and LBW.Alabama birth records for 1990–2010 warm season months (May-September) obtained from the Alabama Department of Public Health (ADPH) were available for these analyses. Study protocol was reviewed and approved through the ADPH and University of Alabama at Birmingham Institutional Review Boards. Births were classified as PTB (<37 weeks) and LBW . Births before 24 weeks or less than 200 grams were excluded from analyses ,24. BirtWe merged birth records with Census 2000 ZIP code-level percentages of poverty and self-identified non-Hispanic African American. As has been previously done, we allowed for non-linearity by classifying these variables into quartiles . Since tTo examine long-term yearly time trends in PTB and LBW by category of rurality, we developed logistic regression models both with and without B-splines. Since preliminary models indicated that overall PTB rates peaked in 2005 and LBW rates peaked in 2006, we tested whether linear changes in PTB or LBW rates before and after these respective peaks differed by rurality category using models with year*rurality interaction terms. Separate models used B-splines with odds ratios (ORs) and 95% confidence intervals (CIs) comparing the different rurality categories at three different time points on the splines to determine differences in the likelihoods of PTB and LBW. The three time points selected for each outcome correspond to the first (1991) and last (2010) years in the study, as well as the year corresponding to peak PTB or LBW rates. Crude models were run, as well as models including the individual-level variables of parity (first or subsequent birth), payment method , years of education , race  and a natural spline for mother’s age.To examine relationships between adverse birth outcomes and individual and ZIP code-level poverty and race, mixed-effects logistic regression models were run, which included a random ZIP code intercept to estimate area-level variation ,19,25. AStatistical analyses and figures were produced in SAS v9.3  and ArcMap 10.0 .Table Maps of ZIP code-level characteristics show that percent poverty and African American distributions follow similar spatial trends across Alabama  and urban  ZIP codes still had significantly higher odds of PTB in 2010, relative to large town ZIP codes . In addition, the 4th quartile of percent African American in RUCA-defined small town and large town strata had no births in the 1st or 2nd quartile of percent poverty , so models examining RUCA-specific estimates were not further analyzed. In PTB models, population density significantly interacted with both ZIP code-level poverty and African American in all year-adjusted models Table . Model 1Model 1 relationships between area-level variables carried associations of larger magnitudes with LBW, compared to PTB Table . For exaThe first part of our analysis found that RUCA codes defined isolated rural regions whereas population density metrics allowed for examination of highly urbanized areas. We found that long-term birth trends differed by rurality definition. Due to previously increasing national PTB rates, a concerted effort was initiated in 2006 to reduce PTB  and overChanging demographics and other individual-level characteristics over this time period partially explained the differing time trends, although isolated rural areas maintained elevated adverse birth outcome odds even after these adjustments. Rural regions captured by RUCAs have low commuting flows to metropolitan and micropolitan regions, indicating likely economic and social isolation. Previous literature has found that rural regions have increased prevalence of maternal smoking, and it has been found that lack of social capital is especially tied to maternal smoking in rural regions ,26. RuraOur analysis found that birth outcome disparities attributable to living in low-income African-American communities were heightened in population dense urban areas, compared to less-dense areas. This result is consistent with a previous study in Pennsylvania, where urban areas had higher preterm birth rates, even after controlling for race and income . Much, bAlternatively, it has been previously posited that significant area-level effects may be found not due to causal relationships, but due to structural and residual confounding. Structural confounding may arise from socioeconomic and racial segregation, which in turn may lead to a lack of adequate distributions of individual-level factors within area-level strata ,40. ThisThere are limitations to the current study. Additional individual-level variables, such as smoking status, alcohol use, psychosocial factors, and disease states of the mothers, would be useful in further describing the relationships we found. Another limitation of this study was that we only had availability to warm-season months (May-September). It is possible that time trends and other relationships investigated differ by warm and cool seasons, limiting the generalizability of our results. Our current analysis does not account for area-level variable changes over the 20 year period. It also has been shown previously that ZIP code-level estimates provide less robust effect estimates than census tract level estimates  and thatThis study finds that adverse birth outcome rates remain higher in isolated rural and more population dense areas, and that these disparities are being maintained or increasing over time. This study also found that relationships between ZIP code-level poverty and adverse birth outcomes were significant in urban areas, even after accounting for individual-level risk factors. Interventions to reduce adverse birth outcomes among low SES populations have been found to be effective, particularly among rural populations, and can be implemented to reduce disparities .LBW: Low birth weight; GVF: Green vegetation factor; MODIS: Moderate imaging resolution spectrometer; OR: Odds ratio; PTB: Preterm birth; RUCA: Rural–urban commuting area code; SES: Socioeconomic status; US: United States (of America).The authors declare that they have no competing interests.STK participated in study design, performed the analysis, and drafted the manuscript. LAM contributed to study design and analysis and manuscript editing. BFZ contributed remote sensing expertise and data and manuscript editing. JMG participated in study design, analysis consultation, environmental health expertise, and helped draft the manuscript. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2393/13/129/prepubA) Percent Poverty and (B) Percent African American Quartiles Across Alabama; Spatial Distributions of (A) Preterm Birth and (B) Low Birth Weight Rates Across Alabama.Spatial Distributions of (Click here for fileCrude B-Spline Odds Ratios and 95% Confidence Interval for Relationships Between Rurality and 1991-2010 Alabama Adverse Birth Outcomes, by Year.Click here for fileCovariate-Adjusted B-Spline Odds Ratios and 95% Confidence Interval for Relationships Between Rurality and 1991-2010 Alabama Adverse Birth Outcomes, by Year.Click here for fileDistributions of Alabama ZIP Code Level Percent African American by Percent Poverty Quartiles, Stratified by RUCA 2.0 Defined Rurality.Click here for file


Preeclampsia is a major cause of perinatal mortality and disease affecting 5–10% of all pregnancies worldwide, but its etiology remains poorly understood despite considerable research effort. Parent-offspring conflict theory suggests that such hypertensive disorders of pregnancy may have evolved through the ability of fetal genes to increase maternal blood pressure as this enhances general nutrient supply. However, such mechanisms for inducing hypertension in pregnancy would need to incur sufficient offspring health benefits to compensate for the obvious risks for maternal and fetal health towards the end of pregnancy in order to explain why these disorders have not been removed by natural selection in our hunter-gatherer ancestors. We analyzed >750,000 live births in the Danish National Patient Registry and all registered medical diagnoses for up to 30 years after birth. We show that offspring exposed to pregnancy-induced hypertension (PIH) in trimester 1 had significantly reduced overall later-life disease risks, but increased risks when PIH exposure started or developed as preeclampsia in later trimesters. Similar patterns were found for first-year mortality. These results suggest that early PIH leading to improved postpartum survival and health represents a balanced compromise between the reproductive interests of parents and offspring, whereas later onset of PIH may reflect an unbalanced parent-offspring conflict at the detriment of maternal and offspring health. Relative to the great apes, humans have long pregnancies and relatively high perinatal risks for mother and offspring. This condition is generally perceived as being due to selection for large fetal brains and constraints on evolutionary responses in the pelvic bones of adult females because of trade-offs with efficient bipedal locomotion While it seems relatively straightforward to explain the narrow human birth canal as an unavoidable evolutionary constraint, this kind of interpretation seems invalid for hypertension-induced problems during pregnancy. It would seem incomprehensible that the fetus harms its pregnant mother while being so intimately and obligatorily dependent on maternal health, both before and after birth. And even if any genetic tendencies for fetuses to increase maternal blood pressure would have arisen during our evolutionary past, it seems puzzling why natural selection has not removed such fitness-reducing traits in our hunter-gatherer ancestors. The alternative explanation that preeclampsia and other hypertension-related pregnancy complications are mostly a modern affluent society problem does not hold either, because preeclampsia is also prevalent in developing countries where there is little access to modern health care Parent-offspring conflict (POC) theory General explanations for hypertensive disorders of pregnancy associated with POC would be most compelling if it could be shown that milder forms such as pregnancy-induced hypertension (PIH) are beneficial for offspring health, as that would imply that the overall benefits of hypertension during pregnancy may balance or even exceed the negative costs of the more severe forms such as preeclampsia, rather than preeclampsia imposing a population-level genetic load. Several studies have demonstrated negative effects of hypertensive disorders on developing fetuses and newborns n = 10 and n = 644) or only occurred  in trimester 1, whereas the ‘PIH in trimester 2′ dummy variable  included PIH diagnoses that started , only occurred  or continued  into trimester 2. This “overlapping” design allowed risk ratio comparisons between trimesters, while also adjusting for the effects of PIH that continued across multiple trimesters. This would not have been possible if we had only included the specific diagnoses for each of the trimesters . For PIH in trimester 2, the risk ratio was <1 indicating a lower risk of infant mortality. This difference was not significant when compared to normotensive pregnancies in the same trimester . All infants of the 678 mothers with PIH in trimester 1 survived the first year of life. This zero value for the risk of infant mortality precluded that formal Cox regression coefficients  could be obtained in comparison with normotensive pregnancies, but suggests that PIH-induced mortality is extremely low when it arises early in pregnancy. Mortality beyond the first year of life was not included as most infant mortality was observed during the perinatal period  and respiratory  diseases  confirmP<0.001, P<0.001 in comparison with the PIH cohort). While these birth weight differences are of some interest in their own right Our results indicate significantly reduced first year mortality and later-life disease risks for offspring born to mothers with PIH early in pregnancy, whereas PIH in later trimesters increases these risks, so they come to approach those connected with preeclampsia when PIH occurs in trimester 3. A proximate explanation for this shift may be that fetuses that became endowed with slightly insufficient placentas have ways to counter the threat of remaining undernourished by increasing maternal blood pressure to drive more nutrients across the fetal-placental barrier Another possible proximate explanation for the observed shift in decreased to increased offspring disease and mortality risk is that the mechanisms  underlying early vs. late occurring maternal PIH differ. For example, Yuan et al. From an evolutionary perspective, the original hypothesis by Haig If genetic variation related to POC In conclusion, our study provides epidemiological evidence of health benefits provided by the milder form of a condition that is generally considered maladaptive, and offers a clearer picture of why hypertensive disorders in pregnancy may continue to persist in modern humans. The results of our study confirm that large public health databases are useful for testing questions inspired by evolutionary theory http://www.medinfo.dk/sks). Unique personal identification numbers  were used to link the DNP data with a number of other population registers including the Danish Civil Registration System, the Fertility Database and the Cause of Death registry. We obtained approval for our study from the Danish Data Protection Agency (www.datatilsynet.dk), the Danish National Board of Health (www.sst.dk) and Statistics Denmark (www.dst.dk), where the latter was the direct data provider. All data were previously de-identified and hosted on a secure computer by Statistics Denmark.The ongoing Danish National Patient Registry (DNP) has collected and electronically stored all diseases diagnosed within hospitals in the Danish population since 1977. From 1977–1993 all diagnoses were coded by the International Classification of Diseases, eighth revision (ICD-8) and from 1994 onwards by the tenth revision (ICD-10)  and excluded twin and higher order births . The sample was further reduced to 413,594 after excluding mothers due to: missing values for gestation length in 1977–1978, mismatches between maternal birth-related diagnoses and corresponding births due to short interbirth intervals, birth information erroneously matching with several biological mothers, and having a previous diagnosis of potential impact in PIH or preeclampsia: hypertension, diabetes, hypotension, purpura (purple discolorations under the skin due to inflammation), some circulatory diseases  and some diseases of the genitourinary system and liver . The final sample , or with preeclampsia . The remaining cases were normotensive singleton pregnancies .We extracted information on all mothers giving birth to singletons from 1 January 1977 to 31 December 2007 (Of the women defined as having pregnancy-induced hypertension (PIH) we included only those with the specific ICD-10 diagnoses DO13 and DO139  and about six meetings with her midwife. At each meeting her blood pressure is measured and urine samples analysed (with urine dip stick). Since PIH is one of the potential precursors of preeclampsia (but not necessary leading to it), women with PIH are subsequently monitored more frequently by their GP or midwife (except when PIH disappears). This could explain why there is a higher incidence of women with only a few days hypertension and fewer cases with a consistent PIH diagnose. PIH is registered but rarely treated, which suggests that not all women with gestational hypertension are recorded with this diagnosis at the hospital, leading to an unknown amount of under-reporting for PIH. The specificity of the Danish National Patient Registry has been validated as being high in general, but a control study indicated that PIH is indeed underestimated To assess the risk of PIH on subsequent health of offspring, we created three dummy variables that marked whether PIH occurred in trimester one, two and three  and usedWomen were diagnosed with different degrees of preeclampsia  that incwww.medinfo.dk). These 14 classes capture essentially all diseases that hospitals in Denmark diagnose except for accidents, which we excluded because no relationships with conditions in the womb were to be expected.Using the PNR (CPR) numbers from the 758,524 offspring born to 413,594 mothers, we extracted information on offspring mortality (deaths recorded within one year after birth) and diseases these offspring were diagnosed with from birth (years ranged from 1979–2007) until 31 December 2009. Offspring diseases were grouped according to 14 standard disease classifications  that theWe used Cox proportional- hazard regression for survival data in R version 2.13.2 to estimate risk of mortality (up to 1 year after birth) and disease , average parental age at birth, combined family income, mothers education level. Family income was calculated as the average of general income before taxes (across the years 1979–2007) between the mother and father . Maternal education was recorded in years, beginning from year 1980. We also included the following covariates to adjust for possible confounding offspring effects: sex of the baby, birth weight and length , year of birth in five-year cohorts, and Apgar5 score. Apgar5 is a general health evaluation made by the midwife or doctor of the neonate’s skin colour, breathing, reflexes, heartbeat and muscle tone five minutes after birth. Each factor can give two points, so the score is between 0–10, where 10 is given to the most healthy appearing babies. To account for possible parabolic effects, quadratic terms for maternal and paternal age were included, but subsequently removed as these were non-significant in all analyses. Scatterplots of the different variables were used to visualize and exclude any outliers in the dataset. The cut-offs for gestational length were set to a minimum of 140 days (20 weeks) and a maximum of 315 days (45 weeks). The cut-offs for birth weight were set to a minimum of 200 grams and a maximum of 6,500 grams.P values were Bonferroni corrected.Model diagnostic procedures were run for the Cox regressions including checking for violation of the assumption of proportional hazards, for disproportional data, and for nonlinearity in the relationship between the log hazard and the covariates. All are standard tests used to check whether a fitted Cox regression adequately describes the data. Tests for proportional hazards were based on scaled Schoenfeld residuals and indicated no violation of proportional hazards for any of the covariates. Potential outliers were removed after identifying them by comparing dfbeta values to the regression coefficients obtained from the Cox regressions. No covariates displayed significant nonlinearity as confirmed by plotting martingale residuals against covariates from the Cox regression. All significant Table S1P<0.05, **P<0.01, ***P<0.001. Bolded RR and P values indicate significance after Bonferroni correction (α = 0.05/14 = 0.0035 ). ns P values were also obtained using a resampling procedure in which disease scores were randomly shuffled across individuals in the dataset to obtain a null distribution for each RR where there is no relationship between traits and disease. The P values of these analyses refer to the number of times (out of 5000 permutations) in which the RR pseudo-estimate was equal to or less than the originally estimated RR, with ns indicating significant P values obtained from Cox regressions becoming non-significant when estimated from the resampling procedure.Risk of disease (from birth up to 27 years of age) within the 14 main disease groups depending on whether offspring were born to mothers with PIH  or preeclampsia (after week 20). Values are Risk Ratios (RR <1 when risk is reduced and RR >1 when risk is increased) from Cox regressions, including their 95% confidence intervals in brackets. *(DOC)Click here for additional data file.Table S2List of the maternal pregnancy-related complications (and their ICD codes) used in our study. We grouped the various forms of preeclampsia (# 2–5 below) into one category for results presented in (DOCX)Click here for additional data file.Table S3only a single trimester 1, 2 or 3 .Risk of disease (from birth up to 27 years of age) within the 14 main disease groups depending on whether offspring were born to mothers with PIH diagnosed in (DOCX)Click here for additional data file.


Despite this investment in birth weight, a known survival indicator, total pre-weaning male mortality was statistically significantly higher than female mortality (12% vs. 7% respectively z = 2.06 P = 0.040). Males tended to suffer from crushing by the sow more than females and statistically significantly more males died from disease-related causes. Although males were born on average heavier, with higher body mass index and ponderal index, these differences were not sustained. In addition male piglets showed impaired thermoregulation compared to females. These results suggest male-biased mortality exists despite greater initial maternal investment, and therefore reflects the greater susceptibility of this sex to causal mortality factors. Life history strategies are being displayed by a domestic population of pigs with sows in this study displaying a form of parental optimism by allocating greater resources at birth to males and providing an over-supply of this more vulnerable sex in expectation of sex-biased mortality.For the most part solutions to farm animal welfare issues, such as piglet mortality, are likely to lie within the scientific disciplines of environmental design and genetic selection, however understanding the ecological basis of some of the complex dynamics observed between parent and offspring could make a valuable contribution. One interesting, and often discussed, aspect of mortality is the propensity for it to be sex-biased. This study investigated whether known physiological and behavioural indicators of piglet survival differed between the sexes and whether life history strategies  relating to parental investment were being displayed in a domestic population of pigs. Sex ratio ) at birth was 0.54 and sex allocation  was statistically significantly male-biased at 0.55 (t Applying ecological theories to situations concerning domestic livestock is a method of addressing applied biological issues in farm animal welfare Life history theories predict that there will be sex-biased mortality as a result of the differential costs and benefits of raising the two sexes. The two main and applicable theories are: (i) an adaptive manipulation of the sex ratio  by mothers unable to rear successful sons The second theory of males being more susceptible to mortality because of higher energetic demands associated with sexual dimorphism Fernández-Llario et al. Given the theories already discussed regarding sex-biased mortality, it is therefore relevant to determine whether male and female piglets differ in the extent to which they exhibit survival indicators known to be critical to the neonate. To ensure postnatal survival, the piglet must negotiate a challenging extrauterine environment, which includes a 15–20°C drop in temperature, the movements and behaviour of an unpredictable mother This study was reviewed and approved by SAC's Ethical Review Committee .2  was offered at a rate of three kg per day followed by 0.5 kg increments each day until seven kg, and then followed by one kg increments each day up to a maximum of 12 kg until weaning. Sows had ad libitum access to water from a trough. Farrowing was allowed to occur naturally, although staff were present for data collection. Approximately 24 h after piglets were born they underwent scheduled husbandry procedures, including teeth clipping, tail-docking and ear-tagging with an identification number. Cross-fostering according to functional teat number was a husbandry method that could be used if necessary, however no fostering was done during the course of this study as litter size never exceeded functional teat number. Piglets were weaned at no less than 26 days old.The experiment took place on an outdoor pig farm in Aberdeenshire, with sows and piglets managed under typical commercial conditions. A total of 478 progeny from 36 parturitions (known as and hereafter referred to as farrowings) of fifth parity, Landrace×Large White×Duroc sows, were subject to detailed study over a period of four months. All sows were artificially inseminated in a nearby indoor facility. After being inseminated, sows were grouped together (approximately 20 sows per group) in outdoor gestation paddocks. Sows were moved from gestation paddocks to individual farrowing paddocks approximately 10 days before the farrowing due date. The individual farrowing paddocks were approximately 20 m×20 m, separated by electric fences. Farrowing occurred in double-skinned insulated huts, with sloped wooden walls and a floor area of 3.09 m2 position; 6.5 cm from the midline at the level of the last rib) and condition scores , birth order (BO), birth interval (BI)  and cumulative farrowing duration (CFD) (the elapsed time between the birth of the first piglet and each subsequent piglet). Immediately after birth, the vitality (V) of the piglet was visually assessed and scored using a categorical scale:1 =  No movement, no breathing after 15 seconds;2 =  No body or leg movement after 15 seconds, piglet is breathing or attempting to breathe ;3 =  Piglet shows some movement, breathing or attempting to breathe and rights itself onto its sternum within 15 seconds;4 =  Good movement, good breathing, piglet attempts to stand within 15 seconds.3) and body mass index  were calculated for each piglet. Piglet gender was noted and birth rectal temperature was recorded (BT) using a digital thermometer . The piglet then underwent a righting response test (R): the piglet was placed gently on its back onto a concave beanbag, then it was released and its latency to right itself onto its feet was timed. The piglets were given 15 seconds to right themselves, if they failed to do this within 15 seconds a latency of 16 seconds was noted. Once a piglet had undergone these measurements, a birth order number was written on its back and it was returned to the sow at the birth site. Processing each piglet in this way took approximately four minutes. Piglets were processed with care by trained staff to minimise handling stress and any confounding effects in the data.A blood glucose sample was taken at birth (BG) from the umbilical cord as a mixture of venous and arterial blood and was analysed using a blood glucose monitor . Piglets were weighed at birth (BW) and a crown to rump length (CRL) was measured (the supine length of the piglet from the crown of its head to the base of its tail). From these measurements, ponderal index /crown-rump length (m)Piglet behaviour data were collected from video play-back and latencies to key landmark behaviours were recorded from birth, with processing time subtracted from these latencies:Latency to reach the udder (U): time between birth and when contact between the piglet's snout and the sow's udder was made.Latency to reach a functional teat (T): time between birth and when firm contact between the piglet's open mouth and a functional teat (not a dummy or inverted teat) was made.Latency to suckle (S): time between birth and firm and prolonged (over 2 seconds) contact between the piglet's mouth and a functional teat was made. The piglet typically adopted a brace position, rapid mouth and muzzle motions occurred and frantic teat-seeking stopped.Rectal temperatures were recorded at 1 h (T1) and 24 h (T24) after birth. Approximately 24 h after birth, a second blood glucose (G24) sample was taken when the piglet was tail docked; after the tip of the tail was docked this discarded tissue was used to provide a blood sample. Piglets were weighed again at 24 h old (W24) and the percentage weight change over this 24 h period was calculated. The piglets were then weighed again at four days old (W4) and at approximately 28 days old when being weaned (WW).Sex ratios  were analysed using the test of equal proportions based on the Pearson's Chi-square statistic with continuity correction. The same test was used to simultaneously compare within-litter sex ratios to 1∶1 equality. Note that some litters did not reach an adequate size for the test and were ruled out. Sex allocation  was analysed by the t-test after checking for normality. Generalized Linear Mixed Models (GLMMs) were fitted by Laplace approximation for models of non-normal data, with appropriate link functions and error structures depending on the nature of the response variable. Continuous responses were fitted by REML with a normal error structure and an identity link function. In all mixed models litter effects were included as a random factor. Statistically significant terms were determined at the usual level of 0.05. Univariate P-values for the fixed effects were determined by asymptotic normal approximation. All statistical analyses were conducted using R version 2.13.1 .Mortality variables  for each sex were modelled using GLMMs with logit link function and binomial error structure. Analyses were split according to the two distinct mortality types: firstly, numbers of stillborn piglets relative to numbers of surviving piglets (i.e. those that were weaned) were compared. Secondly, numbers of live-born piglets that subsequently died before weaning relative to numbers of surviving piglets were modelled. With the aim of determining survival indicators for each mortality type, candidate variables were grouped into separate blocks . In order to control for potentially confounding sources of variation, GLMMs were fitted on all the variables within a block, also including sex and all two-way interactions. Collinearity diagnostics were carried out based on Spearman correlation and variance inflation index. From these, it was found necessary for BMI to be used as representative of BW, PI and W24 in the piglet traits block; and U and S represented T in the piglet behavioural traits block. The minimal adequate model for each block was sought by model selection based on AIC and likelihood ratio tests.Mixed models were also used to analyse sex differences between significant survival indicators. For ordinal responses such as BO and CS a cumulative logit link function was specified. BI was fitted to a Poisson error structure and a log link function. Overdispersion was accounted for by adding an observation-level random effect. The simpler Generalised Linear Model (GLM) was used to model responses only varying at litter level.2 pre-farrowing) and sex ratios at birth Spearman's rank correlations were used.Weight and thermoregulation over time were modelled by Linear Mixed Models (LMMs) considering sex, weaning survival status, time and their interactions as fixed effects. As random effects we specified a nested structure consisting of litters and piglets within litters allowing for correlation over time. In the thermoregulation case, weight evolution was also included in the model as a covariate by linking BW to both BT and T1, and W24 to T24. To test the relationship between maternal condition . Within-litter sex ratio biases were not statistically significantly different . Males were observed to be born heavier on average than females (1516 g (±SE 24.8) vs. 1468 g (±SE 23.6), t438 = 1.95, P = 0.052) and had statistically significantly higher mean body mass index (20.3 (±SE 0.26) vs. 19.6 (±SE 0.3), t424 = 2.66, P = 0.008) and ponderal index (75.6 (±SE 1.2) vs. 72.5 (±SE 1.1), t424 = 2.11, P = 0.035). There was a male-biased maternal allocation at birth for mean weight ratio (0.55 (±0.01), t35 = 2.51, P = 0.017).Of the 478 piglets born, 256 were male and 222 were female and the overall population sex ratio  was 0.54. In a Chi-square test this ratio was not statistically significantly different from 0.5 . The within-litter weaned sex ratio was also not significantly different to 0.5 .Total preweaning piglet mortality  was 18.8%, similar to commercial figures reported at the time of the study 16–20% . Within Surviving piglets were compared with stillborn piglets  and with364 = 2.99, P = 0.003), with males having higher BMI on average. When comparing sexes in the postnatal model, males had significantly higher mean BMI than females regardless of survival status , died = 19.06 (±SE 0.70); Female survivors = 20.13 (±SE 0.25), died = 17.79 (±SE 0.92), t398 = 2.22, P = 0.027). Body weight continued to be measured over time and weights were statistically significantly different with respect to survival status  and sex , with surviving piglets being heavier than those that died and the male-bias maternal allocation evident in surviving males at birth and 24 h ) and there was no evidence that sex ratio at birth was correlated with maternal condition at farrowing .Maternal condition did not vary considerably (CS = 3.1 (±0.1) and PThe male-biased sex allocation at birth in this study supports the theory first outlined by Clutton-Brock The ability for the neonatal piglet to preserve homeothermy is a known critical postnatal survival indicator 0.75Metabolic demand increases with body size at around body-weightCompromised suckling success could also affect immunity, by delaying gut closure in utero, male piglets occupied more space and were heavier than females when uterine resources were adequate, however males were lighter and more likely to die when resources in utero were limited. In the current study the non-significant male-bias sex ratio was similar to those reported in other work with both domestic  may be a more challenging activity for males rather than females, as transfer from the native litter to a foster litter requires appropriate behaviours to acquire and maintain a new teat. In addition if males are likely to be more immune-compromised than females, as suggested in this study, untimely removal from the native mother's colostrum and milk supply, containing valuable immunoglobulins as well as dam specific lymphocytes to facilitate cell mediated immunity Greater susceptibility to mortality of the male sex under challenging conditions, such as those associated with competition for functional teats at the udder, is evident from this study. Male piglets displayed poorer thermoregulatory abilities and died more from disease-related causes than females. Susceptible males showed a poor growth trajectory and it is likely that a trade-off between the necessary body size needed to maximise future fecundity and the various physiological adaptations, including thermoregulation and immuno-competence, promoting short-term survival occurred. Studying the prevalence of male-biased mortality under different management systems and its ontogeny is an important research goal with implications for strategies to improve piglet survival.