OR) of 1.13, with the 95% confidence interval (95% CI) ranging from 1.01 to 1.27. The OR of bone tumor for an increase of 500 gram of birth weight was 1.01 . Interestingly, individuals with high birth weight had a greater risk for osteosarcoma  than those with normal birth weight. In addition, in the subgroup analysis by geographical region, elevated risk was detected among Europeans . The present meta-analysis supported a positive association between high birth weight and bone tumor risk.There have been several epidemiologic studies on the relationship between high birth weight and the risk for bone tumor in the past decades. However, due to the rarity of bone tumors, the sample size of individual studies was generally too small for reliable conclusions. Therefore, we have performed a meta-analysis to pool all published data on electronic databases with the purpose to clarify the potential relationship. According to the inclusion and exclusion criteria, 18 independent studies with more than 2796 cases were included. As a result, high birth weight was found to increase the risk for bone tumor with an Odds Ratio ( Bone tumors are the fourth leading type of cancer death among individuals under 20 years old, and the five-year survival is 61.8% in Europe and 79.0% in the USA . AccordiIGFs), an altered immune function and imprinting errors, which was found to increase the risk for the occurrence and development of cancer, had been reported to be excessive among individuals with high birth weight [et al. [et al. [Among the reported 20 more various subtypes of bone tumor, osteosarcoma (OS), Ewing sarcoma (EWS) and chondrosarcoma (CS) are the major diagnosed types . It is rh weight ,11,12,13h weight ,16,17,18h weight ,20,21. H [et al. ; this re [et al. . In addi [et al. . The conThus, to provide a theoretical basis for the prevention of bone tumor, a meta-analysis was performed to pool all relevant published data to investigate the relationship between high birth weight and bone tumor risk. To the best of our knowledge, this work was the first meta-analysis on high birth weight in relation to the risk for bone tumor.birth weight, birth size, bone, cancer, tumor, neoplasm, carcinoma and sarcoma, were applied in the literature search in various combinations. Besides, additional checks were performed on reference lists from eligible studies for further relevant articles. The language of the included articles was restricted to English.To identify relevant studies on the relationship between high birth weight and bone tumor risk, an electronic search was performed in Embase, PubMed and Web of Science prior to June 2015. Index terms, including CIs or provided sufficient information to calculate. Studies of improper type, including editorials, letters, reviews and non-human research, were excluded. Our study was conducted according to the guidelines for meta-analysis (PRISMA) [Studies were taken into account when they satisfied the following inclusion criteria: (1) the study design was observational; (2) the study explored the association between high birth weight and the risk for bone tumor; and (3) the study reported relative risks with 95% (PRISMA) .ORs with their 95% CIs for each category of birth weight. In addition, when adjusted risk estimates were provided in the articles, adjusted variables would also be collected for further evaluation. If different results were reported in the same study based on different statistical models, we would extract the data adjusted with more confounders.Data was collected separately by two authors (Songfeng Chen and Lin Yang) on the basis of the guidelines for meta-analysis . When diQuality assessment was performed by two reviewers (Songfeng Chen and Lin Yang) based on the Newcastle-Ottawa Scale  for obseet al. [et al. [et al. [ORs and 95% CIs were introduced as the common measurement for the relationship between high birth weight and bone tumor risk.In this paper, the overall relationship between high birth weight and bone tumor risk were evaluated on the basis of comparisons between the highest category of birth weight and the reference group in the identified articles. Except for the researches by Gelberg et al.   [p values were calculated to be two-sided, with the statistically significant level to be 0.05.Subgroup analysis by geographical region, age, adjustments and cancer type were performed in this meta-analysis. Besides, in order to explore potential origins of heterogeneity, sensitivity analyses were also employed. As for the evaluation of publication bias, it was estimated by funnel plots and the tests of Begg  and Eggeion 2.2) ,29. OtheThe flow chart of the selection and identification process of our analysis was shown in et al. [et al. [General characteristics had been shown in the et al.  and the  [et al.  Thus, thI2 value of 28.6% had demonstrated a moderate between-study heterogeneity. However, a p value for heterogeneity (ph) of 0.125 in Cochran’s Q test indicated that the heterogeneity was acceptable. Thus, a fixed-effect model was applied to evaluate the relationship between high birth weight and bone tumor risk and the forest plots of the results were displayed in p = 0.029).According to the results, et al. [et al. [et al. [et al. [et al. [et al. [et al. [et al. [p value of 0.935 for nonlinearity suggested the linear relationship between birth weight and the risk for bone tumor within the included studies. Thus, a linear model was suggested in this dose-response analysis. The summary OR of bone tumor for an increase of 500 gram of birth weight compared with the reference group was 1.01 .As for the dose-response analysis, a total of eight independent reports, including Schuz et al.  2007), 07, et al [et al.  (1998) a [et al.  (1998) b [et al.  (2006),  [et al.  (2014) a [et al.  (2014) b [et al.  (2014) cOR = 1.13, 95% CI 1.01–1.27; random-effect model, OR = 1.13, 95% CI 0.96–1.33; quality-effect model, OR = 1.13, 95% CI 0.92–1.38). Though the results from these three models were similar, the association did not reach a significant level under the random-effect model and quality-effect model. Secondly, the leave-one-out method was conducted through eliminating one study successively. As shown in CI 0.99–1.27) to 1.15 (95% CI 1.03–1.29). All the results were of significance or marginal significance.Sensitivity analyses were conducted in our meta-analysis. Firstly, various modeling methods were introduced , while a non-significant relationship was determined among the American population . Only a marginal relationship was found in the age groups, both in the teenage group  and mixed age group . Further, it should be noted that heterogeneity was low in the mixed age group . Considering adjustments as an important factor in estimating risks, we also extracted the data of adjustments adjusted to gain ORs and 95% CIs in the identified papers. Included studies were classified into two group. It was found that there was an increased association among the group with less than three adjustments , while a non-significant relationship was detected among the group with three or more adjustments . Subgroup analysis on the type of bone tumor was also performed for further evaluation. The reported data of bone tumor was divided into five groups, including groups of combined bone tumor, OS, EWS, CS and other type of bone tumors. Increased risk was found between high birth weight and bone tumor risk among the groups of combined bone tumor  and OS , while no association among the other three groups reached significant levels . Notably, no heterogeneity was found among the groups of bone tumor , or Ewing sarcoma of bone  and chondrosarcoma , which might partly explain the between-study heterogeneity.To explore the potential between-study heterogeneity and evaluate the effects of various aspects on the association between high birth weight and the risk for developing bone tumor, subgroup analyses of geographical region, age, adjustments and cancer type were conducted. OR = 1.21, 95% CI 1.03–1.42, p = 0.018) and age-adjusted group , while no evidence suggested the potential relationship within the teenage group  and non-age adjusted group . With regard to EWS and the CS group, all reported data was adjusted with age. However, no significant association was indicated between high birth weight and the risks for EWS  and CS  among the all four subgroups.To further evaluate the relationship between high birth weight and the risks for the three main subtypes of bone tumor, OS, EWS and CS, we had conducted an analysis among age-related group. As shown from p = 0.910; Egger, p = 0.964); therefore, no publication bias was confirmed in our analysis.The results of the funnel plot analysis for publication bias, where little asymmetry was found, had been shown in IGF1 levels, sex and growth hormones would reach their maximum during the period of adolescent growth spurt and puberty, which might contribute to the development of bone tumors. As an important birth factor to reflect the in utero growth and development, birth weight is related with subsequent bone development and growth trends in puberty. Several mechanism studies suggested that high birth weight might play an important part in the occurrence and growth of bone tumors [IGFs was suggested to play an important role in cancer development [Since bone tumor incidence tends to increase around puberty, many investigators had suspected that bone tumors might be related with growth and development factors . As repoe tumors ,38,39,40e tumors . This sue tumors ,43. Addielopment ,44,45. Ielopment . In receOR = 1.13, 95% CI, 1.00 to 1.27) possessed a lower risk than the interview/questionnaire group , which might be due to the different recall bias. As for the interview/questionnaire group, interviewees might overstate the actual birth weight, when considering the high birth weight as an increased risk for bone tumor. Besides, recall bias would occur resulting from long time lag between birth time and disease diagnosis. In contrast, the data from birth records or birth registries should have no recall bias; thus, conclusions from the birth record group were assumed to be more reliable. In addition to the analyses on the source of data, selection and investigation bias were also considered using the Newcastle-Ottawa Scale. The estimated score ranged from four to eight, indicating the good quality of the included studies. In addition, the results of publication bias suggested that there was no evidence of publication bias in our analysis, thus the method of pooling published data was reasonable.Meta-analysis has the advantage of clarifying the association with a larger power than that from separate studies. However, potential bias should be taken seriously in the meta-analysis . Within I2 value of 28.6% and ph of 0.125 had indicated that the heterogeneity was moderate but acceptable. To evaluate the potential source of heterogeneity, subgroup analysis on type of cancer was performed. In our meta-analysis, there were two studies on combined bone tumor, six studies on OS, four studies on EWS, three studies on CS and three studies on other types of bone tumor. As seen in OR = 1.49, 95% CI 1.03–2.13) and OS group , while no increased risk for bone tumor was detected among the EWS group, CS group and other type group. The significant and positive relationship between high birth weight and OS risk might be ascribed to the following reasons. On one side, birth weight could reflect intrauterine growth and development and might predict a pubertal growth spurt or rapid bone growth during adolescence [IGF-I was suggested [IGF-I. IGF-I was involved in normal bone growth and differentiation, and the association between the growth of OS cell and the modulation of IGF-I is reported in many molecular biology experiments [IGF-I and other growth factors might be ascribed to the findings in our study. Previously, a pooled analysis on the association between birth weight and OS risk with 434 cases was conducted by Mirabello et al. [OR 1.35, 95% CI 1.01–1.79). In our analysis on the OS group, there were 1571 cases included and the results in our analysis was in agreement with the previous research, further demonstrating the positive association between high birth weight and OS risk. However, the potential relationship among the EWS group and CS group should not be denied because the sample size was much smaller than that of OS group. For CS group, since incidence of CS rose with age, and occurred rarely in childhood, there were only 123 cases included to evaluate the association among the CS group. With regard to the EWS group, 937 cases were eligible in our analysis.Though the etiologies of bone tumors are similar, they are different in many aspects ,48. Thuslescence . Birth wlescence ,49. On tuggested ,51. Higheriments ,53. The o et al. . It was OR = 1.05, 95% CI 0.82–1.34) had a lower risk than the group with less adjustments , but failed to reach a significant level. Generally, the results adjusted with more confounders were thought to be more reliable. According to our results, the potential association between high birth weight and the risk for bone tumor among the overall population was not stable and robust. However, we should note that in this subgroup analysis by adjustments, the confounding effect of age, sex and other potential confounders have been mixed together, leading to unexpected outcomes.Bone tumors are reported to be common among childhood and adolescent individuals, and they were more likely to occur among males than females . Thus, aOR = 1.25, 95% CI 0.91–1.72, p = 0.164), while a positive and significant association was detected among the mixed age group . Though the incidence of OS was reported to be more common among younger individuals, the results suggest that the relationship between high birth weight and bone tumor risk was not age-distributed. In addition, the influence of age was further investigated depending on whether age was adjusted or not in the original studies. Consequently, there was a significant and positive association between high birth weight and bone tumor risk after the age at diagnosis was adjusted , while no significant relationship was detected within the non-age adjusted group . The combined results from age-adjusted data were more reliable and convincing. The results in this analysis further confirmed the positive association between high birth weight and the risk for OS. For ES and CS subtype group, all the included studies were adjusted with age, but the results suggest there was no significant relationship confirmed among all the groups, as shown in As for the major confounding factor of age, attention is deserved since various subtypes of bone tumor were related to the different age distribution. It had been reported that Ewing’s sarcoma and OS were the most frequent histologic subtypes in the first two decades, while CS showed an increased incidence after the fourth decade . Particuet al., tumor protein p53 (TP53) germ line mutations was found to be related with OS [IGF had been indicated to be involved in the interaction between high birth weight and bone tumor risk. In our analysis, subgroup by geographical region was conducted, with a significantly increased risk found among the European group  but not the group from America . These results might be due to the different ethnic constitutions of these two groups. It was reported that incidence of bone tumors, especially for EWS and CS, are more common in Caucasians than Blacks [OR of 1.13 for the association between high birth weight and the risk for bone tumor among the overall population. As for the European population, the power was calculated to be 26.8%, while the power among the America group was 18.2%. Therefore, the small statistical power due to the small sample size might contribute to the different findings between Europe and America groups. More relevant studies on American populations are required to increase the power for reliable conclusions.Genetic factors were reported to play an essential part in the occurrence and development of bone tumor. As reported by Savage  with OS ,55. When with OS ,56,57,58 with OS ,60,61. Bn Blacks . Howevern Blacks , the adjn Blacks , the estSeveral limitations should be noted in the present meta-analysis. First, since relevant studies were limited, major confounding factors, including gestation period, dietary factors detailed data on sex and environmental exposure factors, were unable to be studied. Second, only published data was included in our electronic search, which could lead to the potential publication bias. Third, there were only American and European populations included in our meta-analysis, which might prevent our conclusions from extending to Asian and African populations. Relevant studies on Asian and African population are needed for further evaluation.In the present analysis, 18 independent studies with more than 2796 cases were included. The results indicated positive and significant relationships between high birth weight and bone tumor risk. Further, individuals with high birth weight were found to be more likely to develop OS, while European populations with high birth weight exhibited a greater risk for bone tumors. While the sample size of our meta-analysis was sufficient, caution is required due to the inevitable heterogeneity. Additional studies of the biologic mechanisms underlying this relationship are necessary to allow more definitive and reliable conclusions.


Combination antiretroviral therapy (cART) is successfully used for prevention of perinatal HIV transmission. To investigate safety, we compared adverse events (AE) among infants exposed to different maternal cART regimens. We reviewed 158 HIV-uninfected infants born between 1997 and 2009, using logistic regression to model grade ≥1 AE and grade ≥3 AE as a function of maternal cART and confounding variables . Frequently used cART regimens included zidovudine (63%), lamivudine (80%), ritonavir-boosted lopinavir (37%), nelfinavir (26%), and atazanavir (10%). At birth, anemia occurred in 13/140 infants (9%), neutropenia in 27/107 (25%), thrombocytopenia in 5/133 (4%), and liver enzyme elevation in 21/130 (16%). Corresponding rates of AE at 4 weeks were 59/141 (42%), 54/130 (42%), 3/137 (2%), and 3/104 (3%), respectively. Serious AE (grade ≥ 3) exceeded 2% only for neutropenia . Compared with infants exposed to maternal lopinavir/ritonavir, infants exposed to nelfinavir and atazanavir had a 5-fold and 4-fold higher incidence of AE at birth, respectively. In conclusion, hematologic and hepatic AE were frequent, but rarely serious. In this predominantly protease inhibitor-treated population, lopinavir/ritonavir was associated with the lowest rate of infant AE. The prevention of perinatal transmission of HIV is one of the most successful public health interventions of the last few decades. The use of maternal combination antiretroviral therapy (cART), when combined with infant postnatal prophylaxis, has reduced transmission rates to less than 1% in developed countries . This isSpontaneous preterm delivery and low birth weight have been associated with HIV infection during pregnancy. Although not a uniform finding across studies, the use of protease inhibitors (PI) during pregnancy may increase the incidence of preterm delivery –6. MitocAs new ARV and more complex cART regimens are administered for the prevention of perinatal HIV transmission, it is important to identify the toxicities associated with both established and novel regimens in order to inform best choices –31. PrioThis study was approved by the Colorado Multiple Institutional Review Board and exempted from informed consent. This was a retrospective chart review of 190 pregnancies complicated by HIV infection that were managed by the Children's Hospital Immunodeficiency Program (CHIP) in Denver. CHIP is the reference center for the care of HIV-infected pregnant women in Colorado and neighboring states. Data were abstracted for all pregnancies from 1997 to 2009 that resulted in a live infant birth. Maternal data collected included demographics, ARV use, illicit drug use, mode of delivery, hematologic and hepatic laboratory values, CD4 count, and viral load. Undetectable viral load was defined as <400 copies/mL, as this was the lower limit of detection for the earliest data. Duration of maternal viremia during pregnancy was defined as the number of days from the estimated conception date until either the date of the first undetectable viral load measurement after which all subsequent viral load measurements were undetectable or, if there was no sustained viral suppression, the number of days between the estimated conception date and the date of infant birth. The estimated conception date was calculated from infant gestational age determined at birth. Infant data collected included gestational age (preterm defined as <37 weeks), birth weight  defined as <3rd percentile of expected weight for gestational age), postnatal ARV prophylaxis, laboratory values , and hospitalizations or illnesses. Infant complete blood count and liver function panel was assessed at birth (age 0–7 days) and at 4 weeks (±2 weeks). Laboratory toxicities were graded using the Division of AIDS Table for Grading the Severity of Adult and Pediatric Adverse Events . If the Pregnant women received a clinician-prescribed antenatal ARV regimen that consisted of cART (≥3 ARV from ≥2 ARV classes) in most cases; modifications during the pregnancy were based on viral genotype, virologic response, safety, and tolerability, as previously described –35. PI sMost infants were prescribed postnatal prophylaxis consisting of six weeks of ZDV. In some infants, ZDV was replaced by stavudine (d4T) due to anemia, neutropenia, or both. In situations where the risk of perinatal transmission was increased, infants received two- or three-drug ARV. A detailed description of the infant postnatal prophylaxis prescribed and associated adverse events has been previously reported .t-tests, chi-square tests, and Fisher's exact tests, as appropriate. Because most infants were exposed to an antenatal ARV regimen containing two NRTI and a PI, AE were compared between infants exposed to different NRTI and between infants exposed to different PI. The number of infants exposed to nevirapine (n = 10) was too small for comparison. Rates of preterm birth and SGA were compared as a univariate analysis. Logistic regression was used to model grade ≥1 laboratory AE and grade ≥3 laboratory AE (yes/no) as a function of maternal ARV received for ≥28 days during the 35 days immediately prior to delivery and any confounding variable that changed the odds ratio  by >20%. Confounding variables used in the multivariate analyses include preterm birth, infant exposure to ZDV monoprophylaxis versus combination prophylaxis, mechanism of delivery, maternal race and ethnicity, maternal illicit drug use, and duration of maternal viremia during the pregnancy. Statistical significance was defined by p ≤ 0.05.The analysis included infants with at least one laboratory value at either birth or 4 weeks. The statistical analyses require that all observations (infants) are independent of one another or the results may be biased. Given that twins have identical exposure in utero and in order not to underestimate associations between AE and ARV, the twin with the lower grade toxicity was excluded from each pair. Maternal and infant characteristics were compared using 3 in early pregnancy. One hundred twenty-seven women (80%) received cART for at least 28 days prior to delivery. Fifteen women received cART for less than 28 days prior to delivery. Sixteen women received a less intensive ARV regimen. Maternal ARV received in the last month of pregnancy are detailed in http://dx.doi.org/10.1155/2016/9848041).One hundred and sixty-five mothers were managed at CHIP for 190 pregnancies, resulting in 196 live births. Of these, 163 infants (5 twins) had at least one laboratory value available for analysis. The analysis cohort consisted of 158 mother/infant pairs after excluding one infant from each twin pair. There were no perinatal transmissions of HIV. The infant demographics and maternal HIV and obstetrical characteristics are reported in Laboratory AE in the first four weeks of life were common, with three quarters of infants having an AE (any grade) at either birth or 4 weeks. The frequency of an AE grade ≥3 in any laboratory category was 12% at birth and 16% at 4 weeks. Neutropenia was most common, with 25% and 42% of infants having any grade AE, and 13% and 9% having grade ≥3 AE, at birth and 4 weeks, respectively . Anemia p = 0.03) or grade ≥3  at either birth or 4 weeks. Four preterm infants required blood transfusions due to severe anemia. No infants suffered a serious bacterial infection or a bleeding disorder as a complication of neutropenia or thrombocytopenia, and no infants developed liver failure as a result of hepatic inflammation. No infants developed renal or cardiac dysfunction.The 26 (17%) infants born preterm were more likely than term infants to develop an AE of any grade , neutropenia (OR 10.6 (95% CI 1.7–66.4), p = 0.01), and any AE (OR 5.3 (95% CI 1.9–14.9), p = 0.002) (p = 0.046) with the difference primarily due to higher rates of neutropenia (OR 18.3 (95% CI 1.2–267.8), p = 0.03) , = 0.002) . This as= 0.002) . The dif= 0.002) . Exposur = 0.03) . At 4 we = 0.03) .p = 0.05). However, the association of infant AE with NFV versus LPV/r exposure remained significant when duration of maternal viremia was included as a covariate in the multivariate analysis (OR 7.9 (95% CI 2.4–26.2), p = 0.0007). In addition, a univariate analysis demonstrated that the duration of maternal viremia did not correlate with the frequency of infant AE (OR 1.0 (95% CI 0.99–1.0), p = 0.57). An anticipated difference between the treatment groups was the year of delivery. NFV was prescribed in an earlier time period (1997–2007); LPV/r was first prescribed in 2003 and surpassed NFV as the most commonly used PI at our center by 2004. Other variables did not differ including the duration of antenatal exposure to NFV or LPV/r or the proportion of women who received the PI prior to conception (Supplementary Table  1).Additional maternal and infant characteristics were examined to determine whether variables not included in the multivariate analysis might contribute to differences in infant AE at birth for the NFV and LPV/r groups, which had a sample size large enough for analysis (Supplementary Table  1). The mean estimated duration of maternal viremia was longer in the NFV-exposed group than in the LPV/r-exposed group .p = 0.26) or SGA (2/16 (13%), 7/56 (13%), and 4/35 (11%), p > 0.99) between infants exposed to ATV, LPV/r, and NFV, respectively, although the small sample size limited the power to detect a difference.Rates of preterm birth and SGA for infants exposed to different maternal PI were compared in a univariate analysis. No significant differences were found in rates of preterm birth (5/16 (31%), 9/58 (16%), and 5/37 (14%), Low-grade laboratory AE were common at birth and 4 weeks in the cART-exposed infants, but grade ≥3 AE were rare. Anemia and neutropenia made up the majority of AE, and neutropenia comprised nearly all grade ≥3 AE. Our results are in agreement with those of previous studies, which showed frequent, but low-grade, hematologic abnormalities among ARV-exposed infants. Pacheco et al. reported significantly decreased infant hemoglobin and neutrophil values, and Mussi-Pinhata et al. described anemia at hospital discharge in 24% of ARV-exposed infants, although in both studies nearly all AE were of grade 1 or 2 , 23. ReaWe found increased infant laboratory AE at birth associated with exposure to maternal NFV and ATV compared with LPV/r. The difference was observed even when the analysis was restricted to mother/infant pairs with the same NRTI background. At 4 weeks, no difference was detectable between these groups, suggesting that the effect is either transient or masked by the effect of postnatal ARV. Our multivariate analysis controlled for exposure to postnatal ZDV monoprophylaxis versus combination prophylaxis. The association of laboratory AE with particular antenatal ARV within the PI class is reported here for the first time. Of note, Bellon Cano et al. described increased anemia, thrombocytopenia, and liver enzyme elevation after exposure to PI-based cART (of which most regimens contained NFV) compared with two or three NRTI or NNRTI-based cART .The mechanism of the PI effect on hematologic AE is unclear. Fetal hematopoiesis may be more susceptible to the effects of PI, as it occurs mainly in the liver and peripheral lymphoid tissues. PI cross the placenta to some degree; albeit due to their large molecular weight and high protein binding capacity, they do not reach therapeutic levels in cord blood. Cord blood to maternal plasma ratios for ATV, NFV, and LPV range between 0.13–0.24, 0.16–0.22, and 0–0.49, respectively . AmniotiBecause this was a retrospective study, it is possible that the maternal receipt of various PI could be a marker for another confounding variable that contributed to infant AE. Examination for potential confounding variables between the NFV and LPV/r groups identified that on average women receiving NFV had a longer duration of viremia than women receiving LPV/r. However, viremia did not correlate with increased risk of infant AE, and inclusion of maternal viremia as a covariate did not alter the relationship of infant AE with exposure to NFV versus LPV/r.Owing to its limited antiviral activity, NFV is no longer recommended for use in pregnant women . HoweverWe found that ATV exposure was associated with a higher incidence of neutropenia compared with LPV/r. However, infants in this study who were exposed to ATV did not show increased rates of liver enzyme elevation or hyperbilirubinemia compared with infants exposed to other PI. These findings are in contrast to those of Mandelbrot et al. who described a series of twenty-three infants exposed in utero to ATV, nine of whom developed elevated serum bilirubin concentrations at birth and five required phototherapy in the first three days of life . NotablyAlthough ZDV is commonly associated with hematologic side effects in exposed infants, we did not find more hematologic AE at birth among infants exposed to antenatal ZDV compared with those exposed to ABC, d4T, or TDF, but the number of infants exposed to the alternative NRTI was small. Few studies compare infant toxicities after in utero exposure to different NRTI, although Vivanti et al. demonstrated decreased genotoxicity in cord blood of infants exposed in utero to TDF versus ZDV .Preterm birth has been described previously in association with in utero exposure to PI –6. We diWe acknowledge several limitations in this study. The sample size was relatively small. Mothers and infants were not randomized to various ARV in our study; they received a regimen that was clinically appropriate based on severity of disease, tolerability, viral response to therapy, and risk of perinatal HIV transmission. Although two groups of infants for whom we found a significant difference in AE (those exposed to LPV/r and NFV) were cared for in different time periods, we also found differences between the effect of LPV/r and ATV exposure, whose use was roughly contemporaneous. Despite these limitations, this study permits comparison of infant AE within classes of ARV, whereas most similar studies have compared between classes of ARV. It also provides data on newer ARV used for prevention of perinatal transmission.In summary, when used in maternal cART for prevention of perinatal transmission of HIV, NFV and ATV were associated with increased infant AE compared to LPV/r. As ATV use in pregnant women is increasing, larger, preferably randomized studies of infants exposed to LPV/r versus ATV/r are needed to evaluate laboratory AE and to clarify any association with prematurity. Investigations into the safest maternal ARV regimens and timing of ARV initiation are important for optimizing outcomes for HIV-exposed infants, especially those with higher risk of hematologic and hepatic toxicity, such as preterm infants and infants in resource-limited settings.Supplementary Table 1 shows a comparison of maternal and infant characteristics for groups exposed to either lopinavir/ritonavir or nelfinavir.Supplementary Table 2 shows the frequency of infant adverse events at birth associated with exposure to either maternal zidovudine or maternal abacavir, stavudine, or tenofovir.


The notion of a critical community size (CCS), or population size that is likely to result in long-term persistence of a communicable disease, has been developed based on the empirical observations of acute immunizing infections in human populations, and extended for use in wildlife populations. Seasonal birth pulses are frequently observed in wildlife and are expected to impact infection dynamics, yet their effect on pathogen persistence and CCS have not been considered. To investigate this issue theoretically, we use stochastic epidemiological models to ask how host life-history traits and infection parameters interact to determine pathogen persistence within a closed population. We fit seasonal birth pulse models to data from diverse mammalian species in order to identify realistic parameter ranges. When varying the synchrony of the birth pulse with all other parameters being constant, our model predicted that the CCS can vary by more than two orders of magnitude. Tighter birth pulses tended to drive pathogen extinction by creating large amplitude oscillations in prevalence, especially with high demographic turnover and short infectious periods. Parameters affecting the relative timing of the epidemic and birth pulse peaks determined the intensity and direction of the effect of pre-existing immunity in the population on the pathogen's ability to persist beyond the initial epidemic following its introduction. However, even above this threshold epidemics may still fail to occur due to stochastic processes. A classical prediction from branching process theory is that, given R0 > 1, the probability that a single infectious individual in a naive population gives rise to an epidemic is equal to 1 – 1/R0 , and immune, R(0) = p [N(0) − 1], with 0 ≤ p < 1 representing the fraction of the population immune prior to pathogen introduction (due to acquired immunity from a previous outbreak or vaccination).The deterministic model was solved numerically using the deSolve package  in R. GiS, I and R) can take only integer values. Six types of events  occur in continuous time with probabilities proportional to their respective rates in the deterministic model. However, because of the time-dependent birth rate, we decided not to use the exact Gillespie algorithm [δt, the number of events of each type i = {1, … ,6} is drawn from a Poisson distribution with mean iδtr, where ir is the rate of event type i, for example βSI/N for an infection. If more events occur than are feasible , the time step is halved and the new Poisson-distributed random numbers are drawn.The core of our study is based on an event-based, stochastic version of the model, where the three state variables , which is expected to occur with a probability equal to 1/R0 [The stochastic model was implemented in R, using the same range of parameter values and initial conditions as for the deterministic model and explored a full factorial set of model parameters. For each set of parameter values and initial conditions, we ran 1000 simulations for a duration of 10 years. This arbitrary time limit was chosen to allow several seasonal cycles to occur, while keeping in mind that our assumptions of constant parameters and closed populations would not be relevant in nature over long periods of time. We estimated the probability of pathogen extinction as the proportion of simulations that reached the state  to 1/R0 , was rec3.(a)For each of the 18 datasets, we rank our three models using AIC; the results are shown in the electronic supplementary material, appendix 1. The periodic Gaussian function ranked highest with 17 datasets. For the remaining dataset (2), the cosine function is ranked first but the periodic Gaussian function still receives substantial support (ΔAIC = 2.7). The step function (which is the most commonly used one in modelling studies) receives very little support across all but one dataset. Plots of the observed and predicted dynamics show that the periodic Gaussian birth rate generally reproduces the shape of the birth distribution quite well .Figure s range from 2.4 to 227, with a median of 30. To put these values into biological context, electronic supplementary material, figure S3 shows the duration of the birth pulse, defined arbitrarily as the period when 95% of yearly births are predicted to take place, as a function of synchrony s.Across our 18 datasets, the maximum-likelihood estimates of the synchrony parameter (b)m to ensure a stationary population size from year to year, the deterministic population dynamics are given bysK is a normalization factor that depends on s only . Numerical solutions of equation (3.1) show that N(t) follows asymmetrical annual cycles with the peak N(t) occurring after the birth pulse peak. A ‘tighter’ birth pulse  generates greater amplitude of population cycles with a shorter time lag . Increasing the turnover (m) while keeping the average population size constant also results in oscillations of greater amplitude .Using the periodic Gaussian birth rate from equation (2.1) scaled with the turnover rate m, s, ν} parameter space. At the higher end of the relevant parameter region (where higher amplitude oscillations are expected), with a rapid turnover m = 2 yr–1  and a tight birth pulse s = 100 (representing 95% of births occurring within 33 days), average population sizes as low as ν = 100 crash with a frequency of only 1% within 10 years .Stochastic simulations of this simple time-forced birth–death process enable us to assess the probability of a population crash across the {(c)R0, as predicted by branching process theory (see §2b).Using our stochastic SIR model, we assess the ability of the pathogen to invade and persist following the introduction of a single case into a closed population. We distinguish between three categories of extinction occurrences: failure to take off , epidemic burnout following the first wave of infection (with a cut-off time of two years post introduction) and endemic fade-out (after the pathogen has persisted for at least two seasons). We focus on the latter two, having confirmed that failure to take off occurs with a probability of 1/s = 0), the probability of pathogen extinction is strongly influenced by demographic turnover m and recovery rate γ. In general, for a given population size ν and a given basic reproduction ratio R0, the probability of extinction increases with the rate of recovery γ and decreases with the turnover rate m (a). The population size itself has a clear positive effect on persistence. For the sake of clarity, we follow Bartlett [m is higher or the recovery rate γ is lower (a). We then use these simple patterns as a background to study the effect of birth pulse synchrony on pathogen persistence and CCS.Conditional on successful invasion and in the presence of a constant birth rate throughout the year (r rate m a. The poBartlett  and defiis lower a. We thes) concentrates the same number of births over a shorter time period, which amplifies oscillations in the deterministic model and tends to increase the probability of pathogen extinction (b). With more acute infections (i.e. higher recovery rates), a tight birth pulse (say s = 100) can increase the CCS by a factor of 40 compared with a constant birth rate (a). The quantitative effect of s on the CCS is generally weaker in longer-lived host species .All other parameters being fixed, increasing the birth pulse synchrony .A closer look at interactions between parameters and model dynamics reveals an unexpected pattern. In the presence of a marked birth pulse, we observe a non-monotonic effect of the turnover rate sistence . As alresistence . Increassistence , which iφ in equation (2.1)). By default we assumed that the introduction of infection took place when births were at their lowest . However, as shown in figure 2e with an average lifespan of 2 years and an infectious period of one month, a shorter lag between pathogen introduction and the next peak of the birth pulse  can result in a dramatic increase in the CCS. The optimal phase difference for pathogen persistence  varies with combinations of m and γ . Epidemic burnout is most likely (i.e. the deepest post-epidemic trough) when the initial epidemic peak coincides with the birth pulse peak, and therefore is followed by a waning of susceptible individuals entering the population and maximal time until the ‘rescue effect’ of the next birth pulse. However, conditional to persistence beyond this first post-epidemic trough, the relative timing of the birth pulse and introduction of infection had little effect on long-term persistence.An additional factor that can modulate the effect of the birth pulse on the post-epidemic burnout is the timing of pathogen introduction in the seasonal cycle (controlled by the phase parameter p) to simulate the effect of reintroduction of a pathogen after a previous outbreak and extinction, or after an immunization programme. Prior immunity reduces the probability of an outbreak, but has a highly variable effect on the CCS . By reducing the effective reproduction ratio of the pathogen, prior immunity slows down the initial invasion. As a result, the birth pulse occurs earlier in the epidemic cycle, which can either shift the epidemic peak closer to the next birth pulse and rescue the pathogen (c) or, on the contrary, shift the epidemic peak from before the birth pulse to after it, resulting in deeper post-epidemic trough (d). In line with previous points, other parameters that affect the relative timing of the epidemic peak to the peak of the birth pulse  will affect the intensity and direction of the effect of p on CCS .We also considered pre-existing immunity in the population m and s, greater fluctuations in total population size resulted in amplified peaks and troughs of transmission, increasing the likelihood of endemic fade-out .A model with density-dependent transmission gave results similar to the frequency-dependent model over the majority of the parameter space, indicating little overall effect of density-dependent transmission on persistence . However, at extreme values of 4.Growing concern about emerging zoonotic infections has stimulated research effort to model the dynamics of pathogen spillover from wildlife into human populations –33. Howes, for synchrony) across available datasets of birth pulses have not previously been quantified, yet span two orders of magnitude and have significant effects on the infection dynamics.We identified data on seasonal birth rates for a range of mammalian species. Whereas numerous studies report the period over which births take place, few provide the time series of birth numbers required to calculate birth rates for such study, even though appropriate data probably exist in raw form. We assessed alternative mathematical functions for the birth pulse using 18 datasets. Our results suggested that the binary step function used in most published models was not a good representation of real birth pulses and the ‘periodic Gaussian’ function may be preferable for this purpose. Estimates of the key parameter controlling the tightness of the birth pulse : long-lived species with small birth pulses tend to experience a single epidemic, which dies out; by contrast, short-lived species with a higher birth pulse can maintain the pathogen for a few seasons but with a pattern of annual peaks and troughs, which often results in stochastic extinction of the pathogen. Early, post-epidemic fade-outs are also affected by the timing of pathogen introduction relative to the birth pulse, as well as pre-existing immunity in the population (e.g. from a recently extinct outbreak).Our stochastic SIR model with annual birth pulses showed that tighter birth pulses tend to drive pathogen extinction by creating large amplitude oscillations in prevalence. In addition, tighter birth pulses result in the population size being lower for a greater proportion of the year, leading to an increased likelihood of stochastic fade-out. The effect of et al. [et al. [Eptesicus fuscus) played a crucial role in the persistence of rabies virus in that host, but their study only considered large populations, hence offering little insight into CCS. Other ecological factors, such as contacts between multiple host species, have been proposed to contribute to pathogen persistence in wildlife by increasing the effective community size [Empirical estimation of CCS in wildlife remains limited . A few set al. , estimat [et al.  showed tity size ,37.m), such as Townsend's vole , the presence of the birth pulse in M. townsendii increases the CCS for a pathogen with an infectious period of one month (γ = 12) from less than 200 to almost 10 000 individuals . An even greater increase in CCS is expected for pathogens with more acute infectious periods . Second, we consider the grey-headed flying fox , which has a highly synchronous seasonal birth pulse , but a low turnover rate (m = 0.14 [b). Our model predicts that pathogens with an infectious period of less than approximately six weeks (γ ≥ 8) could not persist in a naive population with this turnover rate and degree of synchrony . This raises questions regarding the dynamics of pathogens with short infectious periods, such as Hendra virus (γ ≈ 52) [γ ≈ 20 . This suggests that other factors important in viral persistence in this system are absent from our model; for example, age-structure, metapopulations, multi-host systems, within-host persistence and waning immunity.Apart from contributing to theoretical understanding of viral dynamics and persistence, the notion of the CCS has practical applications in wildlife population management. For example, vaccination is often difficult or impractical in wildlife, and it is often recommended in combination with reduction of population size by culling susceptible animals (discussed in –41). How m ≈ 3.3 ). Even wm = 0.14 ), which (γ ≈ 52)  within p(γ ≈ 52) ) resulteet al. [et al. [et al. [Our model provides theoretical insights into the effect of seasonal birth pulses on pathogen dynamics in wildlife populations and a basis for further extension. For example, one extension would be to consider a metapopulation framework, which would allow recurrent introduction of the pathogen. Interestingly, our model predicts that prior immunity can favour the persistence of some pathogens by dampening the secondary outbreak dynamics, a phenomenon related to the ‘epidemic enhancement’ proposed by Pulliam et al.  and appl [et al. . There i [et al. . A landm [et al.  conclude [et al.  and chyt [et al. . Our modDespite growing interest in the environmental and demographic drivers of pathogen cycles in wildlife , the eff


Preterm and low birthweight children comprise approximately 6% of all live births. It is now a well-known fact that premature children experience many oral complications associated with their preterm births. Prematurely born infants have a short prenatal development period and they are prone to many serious medical problems during the neonatal period, which may affect the development of oral tissues. Adverse perinatal factors, premature birth and exceptional early adaptation to extra-uterine life and functional activity may influence dental occlusal development and symmetry in the jaws. Thus, the goal of the present paper is to elucidate further the effect of preterm birth on the development of the dentition.How to cite this article: Zaidi I, Thayath MN, Singh S, Sinha A. Preterm Birth: A Primary Etiological Factor for Delayed Oral Growth and Development. Int J Clin Pediatr Dent 2015;8(3): 215-219. It is now a well-known fact that premature children experience many oral complications associated with their preterm births. Prematurely born infants have a short prenatal development period that makes them prone to various neonatal complications and developmental problems. According to the World Health Organization definition, a delivery is preterm when it occurs before the 37th completed week of pregnancy.3 but, in 1963, Lubchenco et al reported that birth weight is also determined by the fetal growth rate in addition to gesta-tional age. The incidence of preterm birth varies greatly between populations, being an average in the range 4 to 15%.4 With the changing pattern of lifestyle and increase in urbanization, the frequency of preterm children has increased, but along with that the prognosis of their survival has increased.2In the early years of 20th century, prematurity was defined by birth weight under 2500 gm2 Premature children with very low birth weight (VLBW ≤ 1500 gm) or extremely low birth weight (ELBW ≤ 1000 gm) are at greater risk for short and long-term complications like hyperbilirubinemia, perinatal asphyxia, respiratory, cardiovascular, gastrointestinal, neurological problems and nutritional deficiencies. This may also include disabilities and impediments affecting physical growth and mental development.3A number of etiological factors for premature births exist, many of which are associated with maternal and fetal diseases, but often the causes remains obscure.5The early and long-term effects of premature birth on the physical and psychological growth and development of the child are subjects of considerable current interest. Most studies have indicated that in early childhood the preterm children show significant delay in many areas of physical and psychological growth and development. Although ‘catch-up’ growth has been reported in later childhood, some studies have indicated that long-term delays into adolescence may occur.Like other tissues of the body, the oral structures are also affected by premature birth. All these complications have been reported to be the etiological factors behind the disturbed mineralization in primary teeth . The preTooth enamel is the only hard tissue in the body that is not remodeled. As a result, all of the changes in the structure caused by insults during its development are permanently registered.5 Any alteration during the prenatal, perinatal and postnatal periods, i.e. when the enamel matrix is going through the secretion or maturation phase, can result in enamel defects. As the developing tooth germ is sensitive to a wide range of systemic disturbances, it is unable to recover the damaged enamel that often acts as a repository of information of systemic insults received during development.6 All these factors lead to disturbances in normal calcium metabolism.6Enamel formation of primary teeth begins in 14th week of intrauterine life and continues up to the first year of postnatal life.5 Consequently, ameloblasts and odontoblasts can be affected in the pre- and postnatal periods due to various maternal and child risk factors.7 The clinical expression of these systemic and/or local insults during the enamel matrix formation, mineralization and maturation phases result as qualitative (opacity) or quantitative (hypoplasia) defects8 in enamel structure.6The central hypothesis for developmental defects in preterm and low birthweight infants is altered calcium homeostasis that occurs due to systemic illness, initiating from the time of conception till the end of postnatal periods. Other local effects, such as endotracheal intubation and mechanical ventilation during the postnatal period can also contribute to the condition.9 they also found decrease in enamel thickness in low-birth-weight infants when compared with full-term. The reduced enamel in preterm children is likely to have resulted from both cessation/reduction of ameloblastic activity and the reduced supply of mineral to the developing teeth. The rate of apposition of human enamel has been estimated to be 0.023 mm per day.10Preterm children displayed a variety of changes in the enamel like less enamel thickness, increased roughness, pits, etc. on the surface. In the preterm children, the prenatally formed enamel is the most reduced—at a level of approximately 5 to 13 times the thickness of the enamel of full-term children, which directly reflects the shortened duration in the prenatal stage of enamel formation. The same findings have been shown by Grahnen et al,11 This might be because of the reason that in preterm children, the major part of the enamel is mineralized after birth and may thus be subjected to numerous factors which might disturb the mineralization. The postnatal enamel often had partly a zone of hypomineralized enamel that may be attributed to disturbances in calcium metabolism.8The enamel appears rough, granular, and poorly mineralized because prematurely born infants have a substantial rate of developmental defects of enamel. They also tend to have low calcium stores and disturbed calcium metabolism, with the lowest-birth-weight children most severely affected.7These defects are usually located on the primary teeth undergoing mineralization around the time of the premature birth, i.e. the primary incisors, canines and first molars, although the second primary molars may also be involved. It was previously thought that enamel defects were limited to the primary dentition only, as the permanent teeth have not yet begun their formation at the time of the preterm birth. Other studies, however, have indicated that the effects of birth prematurity may extend into the permanent dentition as well. As the permanent teeth are thought to commence their mineralization a few months after the preterm birth, it was hypothesized that there was persistence of metabolic derangements in the VLBW children well past the neonatal period which affected mineralization of the first few permanent teeth.12Eruption of deciduous teeth and their exfoliation followed by eruption of permanent dentition is an orderly sequential and age-specific event, and is considered as an important milestone during child’s development. Racial, ethnic, sexual and individual factors can influence eruption and are usually considered in determining the standards of normal eruption. Tooth eruption is a complex and tightly regulated process which is divided into five stages namely—pre-eruptive movements, intraosseous stage, mucosal penetration, preocclusal and postocclusal stage. The tooth eruption process is influenced by both genetic and environmental factors acting during odontogenesis. The environmental, prenatal and maternal factors, diseases, nutrition, socioeconomic status and climate, etc. may affect the timing of permanent tooth eruption. The eruption of the deciduous teeth has been thought to be more genetically determined than that of the permanent dentition.13 Factors thought to be related to delayed tooth eruption are short gestational period,1415 low birth weight17 and neonatal factors, including complications of prematurity, systemic disorders, duration of oral intubation, average weight gain/day, etc.18The primary teeth normally develop from mid-gestation until the end of the first year of life. This process may be disrupted in preterm infants by nutritional deficiencies, exposure to certain medications, and traumatic oral manipulations. It has been suggested that nutritional deficiencies in early postnatal life play a role in the development of defective or delayed dentition. The timing of tooth eruption in prematurely born children has been found to be delayed, although it has also been reported that maturation of both the primary and permanent dentition does not differ appreciably between pre- and full-term children.7Changes in dental enamel are one of the most noticeable oral effects of preterm birth, and may classically present as enamel hypoplasia which is defined as a quantitative loss of enamel, or as enamel opacity, which is a qualitative change in the translucency of the enamel. These defects are usually located on the primary teeth which are undergoing mineralization around the time of the premature birth, i.e. the primary incisors, canines and first molars, although the second primary molars may also be involved.8The exact mechanism and etiological factors underlying these defects are not fully understood. There are hypotheses that mineral supply deficiency could be an etiological factor. The possible pathogenesis of the dental defects associated with preterm birth may be related to direct damage to the ameloblasts as in maternal infections . Complications of pregnancy that reduce maternal serum calcium concentrations like maternal toxemia and diabetes, hyperparathyroidism, maternal calcium and vitamin D deficiencies, are often associated with preterm birth. Similarly, neonatal hypocalcemia is often associated with traumatic delivery, cesarean section, birth asphyxia and cerebral injuries (Seow 1986).19 Systemic factors, however, such as metabolic and nutritional disturbances and infections associated with the mineral loss, could cause alterations. These defects are usually located on the primary teeth undergoing mineralization around the time of the premature birth, i.e. the primary incisors, canines and first molars, although the second primary molars may also be involved. It was previously thought that enamel defects were limited to the primary dentition only, as the permanent teeth have not yet begun their formation at the time of the preterm birth. Other studies, however, have indicated that the effects of birth prematurity may extend into the permanent dentition as well. As the permanent teeth are thought to commence their mineralization a few months after the preterm birth, it was hypothesized that there was persistence of metabolic derangements in the VLBW children well past the neonatal period which affected mineralization of the first few permanent teeth. Demarcated opacities in the developing permanent teeth can be a result of long-lasting/sudden insults to the enamel-forming cells ‘ameloblasts’ during the secretory phase (in utero), or severe disturbances during the maturation phase in the first year of life.2021 Prior studies22 have reported that health problems, such as infections and respiratory diseases during the first few months after birth were important risk factors for demarcated opacities in the early erupting permanent teeth.6Various complications associated with prematurity predispose these infants to severe metabolic derangements and hypocalcemia, and these conditions may result in disturbed enamel formation. The problem of deranged calcium metabolism occurs to varying degrees in most premature infants, because two-thirds of the individual’s stores of calcium and phosphorus accumulate during the last trimester of pregnancy and preterm infants miss much of this mineral accretion.23 mouth breathing, oral habits, the mother’s and child’s nutrition and health condition, and other perinatal factors, may influence the dentition during the occlusal development period and the growth of the jaws. Individual occlusal relationships have been reported to indicate a dominance of environment over genetic factors, while some combinations of occlusal traits show noticeable genetic influence.24Both genetics and the environment influence the development of the occlusion. Various environmental factors including—disturbances in general health and growth in childhood, masticatory muscle activity, dietary factorsin utero until approximately 20 years postnatally. Variations exist between individuals in the onset and direction of changes, and in the total increments in arch length, breadth and circumference.25 The changes in maxillary and mandibular arch length are not continuous during the development of the dentition, but occur in the form of growth periods, mainly from 4 to 6 and 10 to 14 years of age. Certain patterns of mean changes in the maxillary and mandibular arch length and breadth are generally associated with eruption of the permanent incisors, canines and premolars, when the changes are greatest.25The growth and development of dentition continues from about the age of 5 weeks 26 Imbalances in the mechanical forces acting on malleable tissues may result in deformations, and gravitational and positioning forces may lead to deformations and deviation of the cranial and facial bones in immature or prematurely born infants.14The mandible is composed of different morphogenetic and functional units, among which the condyle is considered to be a growth zone that is affected by functional alterations. The influence of the functioning of the masticatory muscles has been demonstrated in animal studies, which have shown that increased masticatory function may lead to increased sutural growth and bone apposition, and reduced muscle function to a decrease in bone mineral mass.2728 Large tonsils can reduce the space in the mouth and cause the tongue to be low and pushed forward, pressing against the teeth of the lower jaw.2930 Children with a mouth breathing habit have been shown to have a greater prevalence of class III canine and molar relationships, and greater overjet than others; while tongue thrust is an infantile habit that is apt to lead to an increased prevalence of a class III canine and molar relationship and open bite.31Oral habits, such as thumb sucking, mouth breathing or tongue thrusting, and also oromuscular forces affecting the developing dentition in formative periods, are important as etiological factors for malocclusions.Thus, premature birth and consequential adaptation to extra- intra-uterine life has effect on dental occlusal development, and may interfere with the development of asymmetry and lacerization.Preterm birth is an indicator of enamel defects and is a contributing factor of delayed oral growth and development. Considering the precarious conditions and the risks that these children are subjected to in their growth process, development and adaptation to extra-uterine life, it is necessary to bring very low and extremely low birth weight concerns to the attention of health professionals, who can offer full attention to promote better quality of life.


We examined the associations between socioeconomic trajectories from birth to adulthood and gestational age and birth size in the next generation, using linked data from two population-based birth cohorts carried out in a Brazilian city. By comparing socioeconomic trajectories of mothers and fathers, we attempted to identify-specific effects of maternal and paternal socioeconomic trajectory on offspring birth weight, birth length, head circumference and gestational age at birth.2 population-based birth cohort studies were carried out in 1982 and 2004 in Pelotas (Brazil); 156 mothers and 110 fathers from the earlier cohort had children in 2004. Gestational age and birth length, weight and head circumference were measured. Analyses were carried out separately for mothers and fathers. Mediation analyses assessed the role of birth weight and adult body mass index (BMI).Among mothers, but not for fathers, childhood poverty was strongly associated with smaller size in the next generation (about 400 g in weight and 1.5 cm in height) and shorter gestations (about 2 weeks). Adult poverty did not play a role. For mothers, the associations with gestational age, birth length and weight—but not with head circumference—persisted after adjusting for maternal birth weight and for the height and weight of the grandmother. Maternal birth weight did not mediate the observed associations, but high maternal BMI in adulthood was partly responsible for the association with gestational age.Strong effects of early poverty on gestational age and birth size in the next generation were observed among mothers, but not among fathers. These findings suggest a specific maternal effect of socioeconomic trajectory, and in particular of early poverty on offspring size and duration of pregnancy. Social deprivation in early life is associated with child survival, human capital and adult health.3–6Availability of data from cohort studies, mainly restricted to populations living in high-income countries, has also the investigation of the effects of socioeconomic trajectories—from birth to adulthood—on adult health.5However, we were unable to find any reports on socioeconomic trajectories from birth to adulthood on gestational outcomes in the next generation. Since socioeconomic inequalities are associated with preterm birth and newborn size,2Owing to the possibility of confounding, a comparison of the effects of maternal and paternal socioeconomic trajectories may help assess causality. Specific associations with maternal poverty, which are not also found for paternal poverty, would suggest a causal effect that operates through nutrition or other maternal variables. Data from the 1982 Pelotas (Brazil) birth cohort have shown that maternal birth weight and weight gain during infancy were associated with offspring birth weight, The main objective of this study is to examine the effect of the socioeconomic trajectories from birth to adulthood on gestational age and newborn anthropometry in the following generation, using linked data from two population-based birth cohorts carried out in the same Brazilian city. By comparing socioeconomic trajectories of mothers and fathers, we attempted to identify specific effects of maternal trajectories—with emphasis on early maternal poverty—and compare these to paternal trajectories. We also investigated whether offspring outcomes were mediated by parental birth weight and by adult body mass index (BMI), both of which may be affected by socioeconomic position, and in turn may affect offspring birth weight.In Pelotas, a city located in the south of Brazil, three birth cohort studies are under way. All maternities of the city were visited on a daily basis from 1 January to 31 December 1982, 1993 and 2004 and live births whose mothers lived in the urban area of Pelotas were included in the perinatal studies, which gave origin to the cohorts. Since a number of 1982 cohort individuals were expected to have children in 2004, we linked the two databases using information on names and dates of birth of the 2004 parents. Two hundred and fifty-six children born in 2004 were thus linked to mothers (n=156) and fathers (n=110) from the 1982 cohort; 10 children were born to parents who were both in the cohort. The response rate for initial recruitment was 99.3% in both the 1982 and 2004 cohorts. Further information on these two cohorts is available.17The main socioeconomic variable collected in 1982 was family income expressed in minimum wages; this information was collected as a categorical variable with five groups . The corresponding proportions of the sample in each category were 21.9%; 47.4%; 18.5%; 6.5% and 5.7%. Unfortunately, information on income as a continuous variable was not collected. Owing to the unequal numbers of individuals in each category, it was desirable to classify individuals into tertiles to allow the study of change in income levels since childhood. A principal components analysis was carried out using four variables—delivery payment mode  and mother's schooling, height and skin colour, all of which were strongly related to socioeconomic position. The first component was used to derive a score that was then used to rank individuals within family income groups. Cut-off points were then identified within each category so that three nearly equal-sized groups were formed. To build the tertile-equivalent groups, the 1288 individuals in the lowest family income category were added to the 675 poorest individuals in the second category. The next 1979 individuals in this second category formed the second tertile, while all the remaining individuals formed the last tertile. The three groups did not have exactly the same number of individuals owing to ties in the derived score.9In the 2004–2005 follow-up, total family income was collected as a continuous variable, also recoded in tertiles. Tertiles were calculated for the whole cohorts, not only for those who had children in 2004. A new variable was created—socioeconomic trajectory—by combining these two. The nine possible combinations of tertiles in 1982 and 2004 were recoded into four groups: always poor (poorest tertile in 1982 and 2004); never poor (second or third tertile on both occasions); poor→non-poor (lowest tertile in 1982 and second or third tertile in 2004); and non-poor→poor (second or third tertile in 1982 and poorest tertile in 2004).9Newborns from the 2004 birth cohort were weighed by maternity staff at the time of birth using paediatric scales with a precision of 10 g, regularly calibrated on a weekly basis by cohort study staff using standard weights. Length and head circumference at birth were obtained by the interviewers trained for participation in the study using standardised procedure, within 24 h of delivery. An infantometer and an inelastic tape, both with a 1 mm precision, were used to measure length and head circumference, respectively.Gestational age was estimated according to the following order of preference: (A) the date of the last menstrual period ; (B) ultrasonographic evaluation performed before week 20 of pregnancy; or (C) physical examination using the Dubowitz method.Socioeconomic variables (schooling and income) are described for members of the 1982 birth cohort who were mothers/fathers of children who belonged to the 2004 birth cohort. Anthropometric measurements and gestational age of children born in 2004 are presented by means, SD, minimum and maximum values. Means and their respective CIs were calculated for each group of social mobility and analysis of variance was used in the analyses. The effect of social mobility on anthropometric variables and gestational age on the second generation was investigated through linear regression. Adjusted analyses were conducted in three models. Anthropometric variables  of the grandmother were included in the first model. The second model included all variables of the first model plus parental birth weight. Parental BMI collected in 2004 was added to model 2 and were analysed in the third model.Regarding mediation analyses, the total direct and indirect effects were estimated using G-computation.The Medical Ethics Committee from the Federal University of Pelotas approved the cohort studies. In 1982, mothers provided oral informed consent for participation in the study. In 2004, written consent was obtained from the 1982 birth cohort members and from all mothers whose children belonged to the 2004 birth cohort.supplementary table S1. Five women and five men had no information on family income in 2004, and were therefore excluded from the analyses. The mothers of the 1982 children had considerably less education than their sons and daughters. Mothers and fathers came predominantly from the poorest and middle tertiles of income in 1982, because tertiles were based on the whole sample and not only on participants who had children. In terms of tertiles in 2004, the concentration among the poor is even more marked.The characteristics of the 266 mothers and fathers from the 1982 cohort who had children in 2004 are shown in Those who were poor in 1982 tended to continue in the same category up to 2004. Among those who were non-poor (upper two tertiles) in 1982, almost half of those who had children had become poor by 2004.Mean values for the perinatal outcomes among the 256 children included in the analyses are presented in The crude and adjusted regression analyses on the effects of maternal and paternal social mobility on next generation outcomes are shown in supplementary table S2.The major strength of this study is the linkage of data from two prospective, population-based cohort studies. Since both cohorts included all births in the city of Pelotas in a calendar year, it was possible to identify a substantial number of births. In 2005, 75% of the 1982 cohort individuals were traced during a wave of data collection, so that the vast majority of the cohort was still resident in the city. Owing to the higher fertility and younger childbearing age, participants from the 1982 cohort who were born to low-income families were more likely to have had a child in 2004. This applied to mothers and fathers. A limitation of the present analyses, therefore, was that only births to relatively young parents (aged 21–21 years) were studied; it will be interesting to repeat the analyses when data from the ongoing 2015 Pelotas birth cohort become available.The strengths of our study include its prospective design, population-based samples and the fact that measures were taken directly by the research team using standardised methods, rather than obtained through recall or from records.There are several reports in the literature on the association between birth weight in successive generations.The initial focus of the analyses was on socioeconomic trajectories, but our findings show that early maternal poverty has a lasting effect, regardless of whether or not the woman remained poor. Poverty in childhood, but not at an adult age, is associated with sizeable reductions in offspring birth weight, birth length, head circumference and gestational age. Gavin had previously reported a similar pattern of association with poverty school age (11–13 years), but not adulthood, and offspring birth weight. In contrast, we found no association with paternal poverty or socioeconomic trajectory. Associations between socioeconomic factors and offspring birth weight that were specific for mothers, but not for fathers, had been previously reported.20The specificity of the finding for mothers, rather than fathers, suggests that this association is not due to confounding but has a biological mechanism, possibly through maternal nutrition in early life. The importance of early determinants in offspring birth weight is confirmed by other studies. Previous analyses of five cohort studies from low-income and middle-income countries, including the Pelotas cohort, show that mothers who were stunted at the age of 2 years had offspring whose birth weight was on average 80 g lighter than children born to mothers who were not stunted in early life.5The reported effects for birth weight were also observed for birth length, including the specificity of maternal rather than paternal socioeconomic trajectories. Gestational age was also associated with maternal trajectories in a similar fashion, with early poverty resulting in pregnancies that were about 2 weeks shorter in the adjusted analyses, a sizeable effect. We failed to identify any earlier reports on this association. However, head circumference did not seem to be associated with poverty of the mother or of the father, possibly an indication of a brain-sparing effect in which other parts of the fetal body are sacrificed on behalf of preserving brain tissue.22Our analyses suggest that maternal birth weight did not mediate the effect of early poverty on offspring birth weight and birth length, as indirect effects were small. For gestational age, maternal BMI at adulthood captured about one-fourth of the total effect. Poor women are more likely to be overweight or obese in our cohort, Early exposure to poverty among mothers affects the birth size and gestational age of their offspring. Limited evidence is available for the effects of paternal poverty on health and nutrition of the next generation. The current socioeconomic position of the family is an important determinant of size at birth, and to a lesser extent of preterm delivery.Early but not current maternal poverty is associated with offspring birth weight, birth length and gestational age in a prospective two general study in Brazil. These variables were not associated with paternal poverty. The effect of early poverty on birth weight and birth length was not mediated either through maternal birth weight or current body mass index (BMI). Our findings support the concept that exposure to adverse socioeconomic conditions during the first 1000 days, from conception to the age of two years, can affect the nutrition of the next generation.Our findings support the concept of the 1000 days window of opportunity, from conception to the age of 2 years, as essential not only to adult health and human capital


Background. One of the key strategies to reduce fertility and promote the health status of mothers and their children is adhering to optimal birth spacing. However, women still have shorter birth intervals and studies addressing their determinants were scarce. The objective of this study, therefore, was to assess determinants of birth interval among women who had at least two consecutive live births. Methods. Case control study was conducted from February to April 2014. Cases were women with short birth intervals (<3 years), whereas controls were women having history of optimal birth intervals (3 to 5 years). Bivariate and multivariable analyses were performed. Result. Having no formal education , duration of breast feeding for less than 24 months , preceding child being female , modern contraceptive use , and poor wealth index  of respondents were independent predictors of short birth interval. Conclusion. In equalities in education, duration of breast feeding, sex of the preceding child, contraceptive method use, and wealth index were markers of unequal distribution of inter birth intervals. Thus, to optimize birth spacing, strategies of providing information, education and communication targeting predictor variables should be improved.  Although rates of population increase are now much slower in the developed world than in the developing countries, the world's population has been growing rapidly in recent decades. Much of this increase comes from high-fertility Africa countries. Sub-Saharan Africa and Asia particularly have population growth rates that are outpacing their economic growth . EthiopiLike many other Sub-Saharan African countries, fertility rate is profoundly higher in Ethiopia 4.8) as compared to the global figure 2.69) .8 as com. While t.69 .8 asBirth interval, defined as the number of months between the birth of the child under study and the immediately preceding birth to the mother, has a critical effect on the population size and the health status of the mother and her child , 7. EvidBeyond the health and survival implications of high levels of closely spaced and unintended births, high fertility rates accelerate population growth and undermining development efforts across all sectors. Closely spaced births have a potentially devastating impact on both the individual and the society. This pattern, combined with high levels of unplanned fertility, makes it difficult for women to become productive members of society, thereby limiting their contribution to economic development .While it is difficult to pinpoint the exact reasons, it has been established that the undesirable consequences of shorter interbirth intervals on perinatal, infant, and child survival and maternal mortality have been mainly attributed to maternal nutritional depletion syndrome .Sibling competition increases in the situation of shorter interbirth intervals. It is argued that when a newborn comes, it is likely that the family will invest more of its limited resources in the form of care to the newborn and the other children are more likely to suffer or merely receive inadequate share of the resources distributed among siblings .On the other hand, optimal birth spacing yields the greatest health, social, and economic benefits for the family . AlthougThe Federal Ministry of Health (FMOH), reproductive health department, and health bureaus of respective regions have made concerted effort to reduce fertility. They have been applying multipronged approaches at local and national levels, fostering an enabling environment for family planning being at the heart of a variety of important strategies to address optimal birth spacing, and decreasing fertility .Women in developing countries, however, still have shorter birth intervals than they would prefer . In EthiIn Ethiopia like many other Sub-Saharan African countries, fertility and maternal and child mortality are still high. Thus, acquiring research based knowledge of these problems is key step in undertaking appropriate intervention.Therefore, the finding of this study would be helpful in reminding local and possibly nationwide policy makers of how fertility and birth spacing situation looks like in the study area and to design appropriate strategies for encouraging greater use of optimal birth spacing and thereby ensuring further declines in fertility and maternal and child mortality. It is also hoped that the result of this research will be an input for regional and district health care planners and program managers in designing site specific and scientifically sound interventions to address the gap in the utilization of family planning and optimal birth spacing.In addition, the result will provide important information to local health care providers, community based health extension workers, civil organizations, and facility managers to design appropriate interventions suitable to their clients at facility and community level which can play an indispensable role in convincing mothers to optimize their birth interval in the district.Finally, it is also expected to provide baseline information for further robust follow-up studies.Birth spacing is an important maternal and child health intervention. Studies have confirmed that healthy pregnancy timing and spacing are important interventions to improve infant, child, and maternal health. Although every pregnancy carries a risk of maternal death or morbidity, some pregnancies are at higher risk than others .Recommendations for birth spacing made by international organizations were based on information that was available several years ago. Previous evidences by the World Health Organization (WHO) and other international organizations recommended that waiting for at least 2-3 years between pregnancies can reduce infant and child mortality and promote maternal health . HoweverAccording to the study done in Saudi Arabia, the mean duration of interbirth interval was reported to be 2.38 years . AnotherSimilar Study from Rufiji, Tanzania, showed that the median interbirth interval was found to be 33.4 months. The same study revealed that 48.4% of the respondents had a birth interval below the WHO recommended minimum duration of 36 months .Community based cross-sectional study conducted in Urban Saudi Arabia showed that the median duration of birth interval was 33.5 months. In this study, about a quarter of women had mean birth interval of <2 years while 35.1% had waited between 2 and 3 years .Similarly, study from Ethiopia indicated that the majority (57%) of women were practicing short birth interval length (<3 years) with the median birth interval of 33 months .Evidences from different studies have stressed that the role of sociodemographic, economic, and birth history was significant in influencing interbirth intervals among child-bearing age mothers , 17, 18.p < 0.05)  for me I stopped using it and give birth at a short period of time” (a 30-year-old husband from short spacing group). Some of the discussants from short spacing group mentioned that husband, parent, and religious influence could also prevent women from using contraception and practicing optimal birth spacing: “fear of being divorced makes women experience short birth spacing. Lack of education prevents me from influencing my husband and my life” (a 29-year-old mother from short spacing group). Few discussants mentioned that if the preceding child is a female they have an intention to give birth frequently until they get a boy.Understanding the determinants of short interbirth interval is critical for many Sub-Saharan African countries like Ethiopia, where perinatal mortality and fertility remain alarmingly high. Thus, this community based case control study identified factors influencing short interbirth interval among mothers who had at least two live births with their last child born within the last five years in Arba Minch Zuria District, Ethiopia.Even though no statistical association was revealed, the majority of the respondents defined optimal birth spacing correctly. Similarly, almost all FGD male and female discussants, irrespective of their level of birth interval know the year limits for optimal birth spacing. However, discussants were unable to differentiate between optimal and long birth intervals.In this study, the odds of having short interbirth interval were higher among mothers who had no formal education as compared to their women counterparts who attended formal education. This finding is consistent with evidences from study conducted in Saudi Arabia, Nepal, Jordan, and Pakistan , 21, 22.Similarly, mothers who breast-fed the preceding child for less than 24 months were more likely to have short interbirth interval than their counter parts of mothers who breast-fed for 24 months and above. This study finding was congruent with evidences from Manipur, Iran, Jordan, Pakistan, Ahvaz (Iran), Egypt, and Nigeria , 27, 28.Wealth index of the mother was also a strong predictor of short interbirth interval. Consistent with evidence from study done in Saudi Arabia , the oddStatistically significant association was also seen between short birth interval and age of the mother. In line with study from northern Iran, Tanzania, Saudi Arabia, Nepal, Denmark, Jordan, and Pakistan , 20–22,  “if the mother does not get sufficient rest after the preceding birth when she can regain important nutrients, and recovers from physical and psychological health problems the health of the mother herself, and her new born child will be endangered. Yeah, the family will not be stable and peaceful because the disturbance of health condition” .Although no statistical association was demonstrated, the majority of the study participants reported that short birth interval jeopardizes the health of the mother and the new born child. Similarly, irrespective of their level of birth spacing, the majority of the discussants from all categories underlined that short interbirth intervals are by no means important for the health of the mother, the new born baby, the father, the rest of the children at home, and the overall quality of life of the family:When interpreting the finding of the present study, the following limitations should be considered. The source of data for this study was based on the self-report of mothers, and no validation of information was made with any objective sources such as health facility cards except for immunization cards of their children. But respondents were critically informed about the importance of giving accurate information by assuring the confidentiality of their responses and it is logical to assume that biases are less likely in birth interval related events as compared to other sensitive issues. Finally, there could be a recall bias since women were asked for information about events that occurred in the distant past though different life events were used to memorize them.Optimal interbirth interval has significant role in reducing fertility and maternal and child mortality. However, birth interval of children varied with different biological and socioeconomic factors of their families. Educational status of the mother, contraceptive utilization, duration of breast-feeding, sex of the preceding child, age during delivery of the last child, and wealth index of respondents were independent predictors of short interbirth interval between the last two live births. Thus, the Ethiopian Ministry of Health together with its stakeholders should strengthen the existing strategies of providing information, education, and communication giving critical attention to women and their husbands to advance their awareness about the importance of modern contraceptive utilization, breast-feeding, and optimal birth spacing. Policy makers should also design new strategies to encourage women to pursue their education to at least primary school level. Moreover, this study considered only a single birth interval and therefore further researches, involving more than two interbirth intervals, should be done to make these findings more informative.As part of the full length research paper the supplementary material encompasses, location Map of Arba Minch district including all localities, full length English and Amharic  version questionnaire for both qualitative and quantitative studies.The questionnaire has eight variable categories including Respondents' Identification, Socio-Demographic Characteristics, Birth History, Knowledge of women about Birth spacing, Breast feeding practice , Knowledge and Practice of Modern Contraceptive use, Attitude of respondents towards birth spacing, fertility and family planning, and Household wealth status of respondents.