Distal radius fracture is a common injury with a variety of operative and non-operative management options. There remains debate as to the optimal treatment for a given patient and fracture. Despite the popularity of volar locking plate fixation, there are few large cohort or long term follow up studies to justify this modality. Our aim was to report the functional outcome of a large number of patients at a significant follow up time after fixation of their distal radius with a volar locking plate.180 patients with 183 fractures and a mean age of 62.4 years were followed up retrospectively at a mean of 30 months (Standard deviation = 10.4). Functional assessment was performed using the Disabilities of the Arm, Shoulder and Hand (DASH) and modified MAYO wrist scores. Statistical analysis was performed to identify possible variables affecting outcome and radiographs were assessed to determine time to fracture union.The median DASH score was 2.3 and median MAYO score was 90 for the whole group. Overall, 133 patients (74%) had a good or excellent DASH and MAYO score. Statistical analysis showed that no specific variable including gender, age, fracture type, post-operative immobilisation or surgeon grade significantly affected outcome. Complications occurred in 27 patients (15%) and in 11 patients were major (6%).This single centre large population series demonstrates good to excellent results in the majority of patients after volar locking plate fixation of the distal radius, with complication rates comparable to other non-operative and operative treatment modalities. On this basis we recommend this mode of fixation for distal radius fractures requiting operative intervention. Fractures of the distal radius are common -3. IncreThe aim of this study was to review a large group of patients treated with volar locking plate fixation of their distal radius fracture and to assess their functional outcome at a significant follow up time. Secondary aims were to establish if any individual factors affected this outcome and compare our results with those of other studies.Our hypothesis was that functional outcome and complication rate after internal fixation of the distal radius with a volar locking plate would be comparable to, or better than other reported treatment modalities.We conducted a retrospective study of patients with fractures of the distal radius managed by internal fixation using the Stryker Matrix Volar Locking Plate (Stryker Leibinger GmbH & Co. Germany) between June 2004 and October 2007 treated in a single centre. Indications for surgery were displaced intra-articular fractures with post reduction articular step of > 2 mm; radial shortening of > 3 mm or > 15 degrees of saggital plane angulation [Using the departmental trauma database, 221 suitable patients were identified for inclusion in the study. One patient was unwilling to participate; six were deceased; 26 were not contactable and eight were unable to participate for medical reasons. The remaining 180 patients with a total of 183 fractures were followed up by means of case note and operative record analysis, radiographic assessment and functional questionnaire.Case notes were used to establish all demographic details including; mechanism of injury; time to surgery; post-operative immobilisation; surgeon grade and intra-operative details. Complications were also recorded and confirmed with the patients.The functional questionnaire was telephone based, and utilised the 'quick' Disabilities of Arm, Shoulder and Hand (Quick DASH) score -11, and Pre-operative radiographs were classified according to the AO-ASIF classification system  and postCategorical variables and baseline demographic data are described using frequencies and percentages. Continuous variables with a symmetric distribution are presented using means and standard deviations (SD) and continuous variables with a skewed distribution are presented using the median and inter-quartile range (IQR). Demographic variables such as age and gender and operation related variables such as surgeon grade, fracture type and time to surgery were considered to establish if any of these factors affected functional outcome. Spearman's rank correlation coefficient was used to assess the association between continuous variables and non-parametric tests  were used to test for differences in outcome between groups. All tests were two-sided with the level of significance set at 0.01 to take account of multiple testing.Surgery was performed under general or regional anaesthesia with use of an arm tourniquet and administration of antibiotics according to local policy. A standard volar approach through the bed of Flexor Carpi Radialis was performed. The Stryker Matrix Smartlock volar locking plate (Stryker Leibinger GmbH & Co. Germany) was used in all cases. This is a low profile, titanium plate incorporating 20 degree variable angle locking for all screws. Skin closure and postoperative immobilisation was according to the operating surgeon's preference. Bone grafting and carpal tunnel decompression were not routinely performed, although carpal tunnel decompression was done at the time of internal fixation in 10 patients at the discretion of the operating surgeon. Reasons for decompression included pre-existing carpal tunnel syndrome (n = 2); pre-operative symptoms of median nerve compression (n = 5) and excessive intra-operative swelling (n = 3). Ulna styloid fractures were not treated operatively in any patients. Distal Radio-Ulnar Joint (DRUJ) instability was documented in only one case and was treated with Kirschner wire stabilisation for six weeks.The median DASH score for all patients was 2.3 (IQR 0-6.4) and the median MAYO score was 90 IQR 75-100). The MAYO and DASH scores were good or excellent in 133 patients (74%), satisfactory in 41 (23%) and poor in six patients (3%). There was a strong relationship between the two scoring systems with virtually no variation in outcomes. Figure 00. The M132 patients were female (73%) and 51 were male (27%). There was little variation in the MAYO score between men and women, however the DASH score for women was more variable than in men. A Mann-Whitney test showed some indication of a higher DASH score in women than in men (p = 0.025), however when taking into account multiple testing using the Bonferroni adjustment, this is was not considered statistically significant.Mean age was 62.4 years . No relationship was seen between age and either the MAYO or DASH score .Mean time to surgery was 8 days . No relationship was seen between time to surgery and either the MAYO or DASH score .Mean time at follow up was 30 months post operatively . In total, 62 patients (34%) were followed up in their second postoperative year; 69 (38%) in year three; 42 (23%) in year four and seven (4%) in year five. Patients did not have significantly different functional outcome according to their year of follow up.The primary surgeon was a consultant in 56 cases (31%) and was a middle grade (trainee registrar or staff grade surgeon) in 127 cases (69%). In total 26 different surgeons performed all operations. Using the Kruskal-Wallis test there was no evidence of a difference in MAYO (p = 0.38) or DASH score (p = 0.73) for the different surgeon grades.Post-operative immobilisation for between four and six weeks was instituted in 76 patients (42%) as opposed to immediate graduated mobilisation in the remaining 107 patients (68%). This was surgeon dependent and there was no significant discrepancy between the two groups in terms of fracture type or patient age. There was no statistical difference between the groups in terms of functional outcome  or in the incidence Chronic Regional Pain Syndrome (CRPS) (p = 0.09).Table 133 patients (74%) had post-operative radiographs available for analysis. This was because there was a change over to electronic imaging in the first year of the study period making it difficult to access all radiograph hard copies.Overall mean time to fracture union was 8.4 weeks (6-28 weeks). Table In total, 27 patients (15%) suffered a postoperative complication. These are outlined in table We present one of the largest single centre case series of this kind, with good or excellent outcome in the majority of patients at 30 months follow up. While there are long term studies looking at the outcomes of patients treated non-operatively , with exOur study is limited by the fact that it is retrospective. This did not allow us to follow up patients at several time points after surgery and correlate progression in functional score with time. Other authors have demonstrated that improvement in functional scoring is particularly applicable to the first year after surgery ,22. GoldWe used the MAYO wrist score alongside the DASH score to assess functional outcome. There was no discrepancy between the two scoring systems in terms of grading outcome. Although the two scoring systems are not directly comparable, the use of two systems reduces the likelihood of errors generated by the use of a single system. We used the quick DASH score which is comprised of 11 questions and has been shown to have similar cross-sectional and test-retest reliability to the traditional 30 point full DASH score -11. AlthWe had expected fracture type, surgeon grade, time to surgery or patient age to be possible factors, which affected functional outcome, however we were not able to demonstrate this. This is actually consistent with the findings of other studies ,22. It iOnly one type volar locking plate was used for all patients in this series as this was the only implant available at our centre during the study period. Subsequent development has lead to the design of many newer plates with a variety of features including different locking mechanisms, material properties and fragment specific designs. The use of newer plates may have changed outcomes in some way, however the basic principles of fracture reduction, stable fixation and respect for the soft tissues remain paramount regardless of implant.Many studies reviewing various methods of fixation look at radiographic parameters that affect outcome, however few if any have looked at surgeon grade or time to surgery as we have. In a large radiographic study, Mackenney et al. showed that age over 80 years; metaphyseal comminution and positive ulna variance were the main predictors of instability. This and poor radio-carpal alignment were shown to be associated with poor outcome . HoweverAt the time of the study and currently, it is not our practice to treat Ulna styloid fractures surgically. This is based on the fact that they have not conclusively been shown to alter functional outcome ,25,26.The only factor that was associated with a poorer functional outcome was the development of a post-operative complication. The 27 patients with a post-operative complication had both DASH and MAYO scores considerably worse than the other patients, and those with what we classed as a major complication (10 patients) had a much worse functional score. The two patients who acutely developed post operative carpal tunnel syndrome were both females under the age of 60 years. Neither had preoperative symptoms. One had a 23-C2 type fracture whilst the other had a 23-A2 fracture. Ring et al. suggested that younger women with significantly translated fractures may be at higher risk of carpal tunnel syndrome however, other authors have not advocated the need for carpal tunnel decompression as a routine in any fracture type, age group or gender . Whether16 patients developed features of CRPS at some point in their follow up. In order to avoid under reporting the incidence of CRPS, we included all patients who had any mention of CRPS type symptoms  in their follow up records. There was no correlation between fracture type or period of immobilisation and the incidence of CRPS, however, the mean age of those with CRPS was younger (58.9 years) than the group as a whole. Also, 13 (81%) of these patients were female. This finding is in keeping with other reported incidences of CRPS after distal radius fracture . We now It is not viable to statistically compare the DASH and MAYO scores of those with a complication and the general cohort, as 'complications' are an outcome in themselves, and these patients were not pre-defined for comparison. However, it can be inferred that those who sustain a complication are likely to have a worse functional outcome. Our overall complication rate is comparable or better than other reported treatment modalities including non-operative treatment ,17, percAs this is a retrospective non-comparative study, we cannot disregard other treatment modalities for distal radius fracture. However, there is a growing body of evidence that supports volar locking plate fixation as the modality of choice when surgery is indicated. We feel that this study which is the largest of this kind supports this trend but do acknowledge that further high quality trials are necessary to better answer this contentious issue.This large series demonstrates good to excellent results in the majority of patients after volar locking plate fixation of the distal radius, with complication rates comparable to other non-operative and operative treatment modalities. Those who suffer a complication have a much worse functional outcome, hence, while we recommend this mode of fixation for distal radius fractures requiring operative intervention, utmost care must be exercised to prevent complications and identify them early.The authors declare that they have no competing interests.JP, AT and KG collected the data and compiled the patient database. They followed up the patients and wrote the paper. LB performed all statistical analysis. DSE & KJN were responsible for the conception of the study and supervision of the project. They proof read and amended the paper as necessary. All authors have read and approved the manuscript.


Little is known about relationships between dietary patterns, n-3 polyunsaturated fatty acids (PUFA) intake and excessive anxiety during pregnancy.To examine whether dietary patterns and n-3 PUFA intake from seafood are associated with high levels of anxiety during pregnancy.n 9,530). Dietary patterns were established from a food frequency questionnaire using principal component analysis. Total intake of n-3 PUFA (grams/week) from seafood was also examined. Symptoms of anxiety were measured at 32 weeks of gestation with the Crown-Crisp Experiential Index; scores ≥9 corresponding to the 85th percentile was defined as high anxiety symptoms. Multivariate logistic regression models were used to estimate the OR and 95% CI, adjusted by socioeconomic and lifestyle variables.Pregnant women enrolled from 1991–1992 in ALSPAC  and the traditional  pattern scores were less likely to report high levels of anxiety symptoms. Women in the highest tertile of the vegetarian pattern score  were more likely to have high levels of anxiety, as well as those with no n-3 PUFA intake from seafood  when compared with those with intake of >1.5 grams/week.The present study provides evidence of a relationship between dietary patterns, fish intake or n-3 PUFA intake from seafood and symptoms of anxiety in pregnancy, and suggests that dietary interventions could be used to reduce high anxiety symptoms during pregnancy. Excessive anxiety in pregnancy has been extensively investigated as a potential maternal risk factor for adverse outcomes for both mothers and their offspring Anxiety has been extensively investigated, but little is known about nutritional variations that might be associated with a higher likelihood of occurrence of this outcome in pregnancy. Essential n-3 polyunsaturated fatty acids (PUFA) are nutritional components obtained exclusively from dietary sources, and fish and seafood are the major sources Epidemiological studies considering diet and common mental disorders have employed factor analysis to derive dietary patterns. Instead of looking at individual nutrients or foods, this method identifies patterns of food intake and examines its effect on prediction of disease risk To our knowledge nutritional benefits of dietary patterns and fish consumption in the UK have not been fully explored in relation to anxiety in pregnancy. ALSPAC had previously identified five dietary patterns in pregnancy. Thus, the aim of this study is to evaluate whether dietary pattern scores, frequency of seafood consumption and n-3 PUFA intake are associated with risk of high levels of anxiety symptoms in pregnancy. We hypothesized that low scores on health-conscious  and/or traditional  dietary patterns would be associated with high anxiety symptoms in pregnancy. Moreover, we expected that a lower intake of n-3 PUFA, consequence of a low frequency of seafood consumption, would be associated with greater risk of high levels of anxiety symptoms in pregnancy.The ethical approval for the study was obtained from the Avon Longitudinal Study of Parents and Children (ALSPAC) Law and Ethics committee (IRB#00003312) and the local research ethics committees . Written informed consent was obtained from all participants in the study.http://www.bristol.ac.uk/alspac). The representativeness of the ALSPAC cohort with respect to the population living in Avon county and the whole of Great Britain in 1991 has been assessed and is reported on the ALSPAC website .ALSPAC was initiated to identify features of the environment that influence the health and development of children and their parents n = 199) were excluded from the present analysis. We only included the women for whom complete data were available on all confounding variables . After exclusion for missing data, the final samples comprised 9,530 subjects .This manuscript uses the information collected on diet and anxiety from maternal self-completed questionnaire at 32 weeks' gestation. Women with multiple births , a validated self-rating inventory relating to free-floating anxiety, depression and somaticism Women completed a self-reported food frequency questionnaire at 32 weeks of gestation which comprised 110 questions enquiring about the frequency of consumption of 43 food groups and food items, and about daily consumption of a further eight basic foods Since seafood was the predominant source of essential n-3 PUFA in the diet of this population at this time, we estimated the total n-3 PUFA intake from seafood only, using the three questions that assessed seafood consumption: “How many times a week nowadays do you eat (a) white fish , (b) dark or oily , or (c) shellfish ?” Each response had 5 predefined categories: never or rarely, once in 2 weeks, 1–3 times per week, 4–7 times per week, and more than once a day. These were converted to weekly frequencies of consumption (portions per week) of 0, 0.5, 2, 5.5, and 10, respectively. Portion sizes and type of fish were based on typical consumption patterns in UK and n-3 PUFA content was calculated using British food composition tables years) or equivalent; high: A-level  or higher)]; Socio-economic: work status , housing tenure ; crowding, persons per room ; Lifestyle: maternal smoking during first trimester of pregnancy , maternal alcohol consumption during first trimester of pregnancy ; Obstetric: parity , previous history of abortion ; previous history of miscarriage ; Chronic stress and life events: the mother's life events in childhood scale , a scale of 44 recent life events , chronic stress as measured by a family adversity index .To build the theoretical model for high levels of anxiety symptoms, the following independent variables were included in the analyses: Socio-demographic: age , education based on the highest educational qualification achieved [low: no more than a vocational qualification; medium: O-level  and 95% confidence intervals (CIs). In order to allow flexibility of variable selection, those with a P value <0.20 in the bivariate model plus total estimated dietary energy intake to control for different energy requirements. Additionally, on a final model we repeated the analysis adjusting for the total n-3 PUFA from seafood. These analyses aim to test whether the possible dietary pattern effect could occur in detriment of n-3 PUFA. Furthermore, we tested if there was an association with total n-3 PUFA intake from seafood, grouped into none; 0.1–0.4 g; 0.5–1.5 g; >1.5 g per week. The group with the highest intake was considered the reference. We performed logistic regression controlling for all confounders identified in the first stage, and for total energy intake. To test the association between the frequency of fish consumption and anxiety, we repeated the regressions with each of the three categories of fish intake provided in the food frequency questionnaire stratified as: never or rarely; once in 2 weeks or 1–3 times per week or more. Finally we analysed within each dietary pattern, for the women eating the different frequency of dark and oily fish, based on the above-mentioned categories, whether fish itself on top of a dietary pattern could modify the likelihood of anxiety occurrence.Each of the dietary pattern scores was categorized into tertiles and the lowest scoring group was considered the reference category. The independent effect of each dietary pattern was tested on the outcome, adjusting for all identified variables with We performed further analyses comparing those women included and those excluded in the current analysis due to any missing data using chi-square test for proportions for all confounding variables. All statistical analyses were conducted with the IBM Statistical Package for Social Science version 18.0.Table S1 shows the distribution among those included and those not included in the current analysis.The variation in the prevalence of high levels of anxiety symptoms with the confounding variables is shown in In multivariate logistic regression, both health-conscious and traditional patterns showed protective associations, which persisted even after n-3 PUFA from seafood had been taken into account. This was, in particular, due to women classified in the highest tertile of the health-conscious pattern  and to a trend for lower odds with high scores on the traditional pattern. Women classified in the middle  and the highest tertile  of the vegetarian pattern were more likely to have high levels of anxiety symptoms in comparison with those classified in the lowest tertile. The processed and confectionary dietary patterns did not show associations with high anxiety after multivariate adjustment Table 2.grams/week   when compared with those with intake of >1.5 grams/week. The likelihood of high anxiety symptoms was also observed for those with n-3 PUFA intake between 0.1–0.4 grams/week  and for those with intake between 0.5–1.509–1.47) Figure 1Table S2).The frequency of dark and oily fish intake among women inside each dietary pattern revealed a significant increase in the odds for high anxiety symptoms when those with intake of 1–3 times a week or more were compared with those that consumed fish rarely or never. The observed odds were as follows: health-conscious ; traditional ; confectionary  and vegetarian . The results were significant among all dietary patterns, except for the processed one (P = 0.06). To our knowledge this is the first study describing a relationship between dietary patterns and high levels of anxiety symptoms during pregnancy, and the first to provide evidence of low n-3 PUFA intake and low fish consumption being associated with high anxiety symptoms or anxiety-related conditions.The main finding of the present study is that women that scored in the highest tertile of health-conscious or traditional dietary patterns were 23% and 16% less likely to have high levels of anxiety symptoms at the 32 weeks of pregnancy, respectively when compared with those women classified in the lowest tertile of such patterns. Women that scored in the highest tertile of a vegetarian pattern had 25% greater likelihood for high levels of anxiety symptoms, when compared with women classified in the lowest tertile. In addition, women with no intake of n-3 PUFA from seafood or that reported never or rarely for frequency of dark or oily fish consumption, had respectively 53% and 38% greater likelihood of high levels of anxiety, when compared with women with n-3 PUFA intake from seafood >1.5 grams/week, or who consumed dark or oily fish 1–3 times a week or more. It was interesting to note that even for women who adhere to a vegetarian pattern, the likelihood of anxiety was 43% higher for those that ate dark and oily fish rarely or never in comparison with those that ate 1–3 times a week or more. We observed a similar behaviour for all other patterns, except for the processed one. The associations remained significant even after adjustment for potential confounders. The dietary patterns analysis results were robust to adjustment for n-3 PUFA intake from seafood. Neither processed nor confectionary patterns were associated with high anxiety symptoms in the fully adjusted analysis, although the later one presented borderline association , which may not reflect the entirety of pregnancy. However, the study of Crozier et al. (2009) Our findings are consistent with previous investigations of dietary patterns and anxiety among adult women, and add new evidence regarding high anxiety during pregnancy. In a cross-sectional analyses of 1,046 Australian adult women followed up in a longitudinal cohort study, Jacka et al. (2010) Our findings of a likelihood of low anxiety symptoms among women with intakes of >1.5 grams/week of n-3 PUFA are in line with previous clinical investigations of PUFA and anxiety disorders. Clinical investigations have provided strong evidence that consumption of long chain n-3 PUFA presents significant anxiolytic benefits. A randomized clinical trial among medical students without a diagnosis of anxiety disorder reported a 20% reduction in anxiety symptoms after 2.5g/day of n-3 PUFA supplementation during a period of stress compared to placebo We observed a protective effect of high scores on traditional and health-conscious dietary patterns and an increased likelihood of high anxiety symptoms with the vegetarian (or vegetable protein) pattern. The main difference between the first two dietary patterns and the vegetarian one is the greater likelihood of animal sources – meat and fish – in the traditional and health-conscious patterns, respectively. It is unlikely that women who were anxious about their pregnancy turned to vegetarianism since in ALSPAC only 4% of vegetarians had been eating in this way for less than one year. In the modern diet, meat is the richest source of vitamin B12, and fish is the main source of long-chain essential n-3 PUFA, especially docosahexaenoic acid . These nutrients are necessary for optimal neurological function and signs of neurocognitive impairment are reflected in mood changes The results further suggest an independent relationship between low n-3 PUFA intake and fish consumption and anxiety. Although no inference can be drawn regarding causation such data do suggest that a lower intake of n-3 PUFA or low dark and oily fish consumption is associated with higher levels of anxiety symptoms. Pregnancy is associated with a significant decrease of n-3 PUFAs in tissues due to high transference to the fetus, which may be related to a decrease in uptake of n-3 PUFA into cell membranes consequent upon fetal demands The present study provides evidence of a significant relationship between dietary patterns and symptoms of anxiety in pregnancy, and suggests that dietary interventions might possibly be used to prevent or reduce high anxiety symptoms during pregnancy. From a clinical perspective, the current findings indicate that dietary counselling during prenatal care with a focus on healthy habits and increased n-3 PUFA intake from seafood consumption may be beneficial to women with high anxiety during pregnancy. However, appropriately designed clinical trials need to be conducted to evaluate the efficacy of such interventions.Table S1Distribution of potential confounders among those women included  and those not included  in the current study.(PDF)Click here for additional data file.Table S2Investigating the frequency of fish consumption within women in the third tertile of adherence to each dietary pattern, a significant increase in the odds for high anxiety symptoms among those women with lower frequency of fish consumption is observed. This trend is found in all dietary patterns, except the processed one.(PDF)Click here for additional data file.


Identifying when past exposure to an infectious disease will protect against newly emerging strains is central to understanding the spread and the severity of epidemics, but the prediction of viral cross-protection remains an important unsolved problem. For foot-and-mouth disease virus (FMDV) research in particular, improved methods for predicting this cross-protection are critical for predicting the severity of outbreaks within endemic settings where multiple serotypes and subtypes commonly co-circulate, as well as for deciding whether appropriate vaccine(s) exist and how much they could mitigate the effects of any outbreak. To identify antigenic relationships and their predictors, we used linear mixed effects models to account for variation in pairwise cross-neutralization titres using only viral sequences and structural data. We identified those substitutions in surface-exposed structural proteins that are correlates of loss of cross-reactivity. These allowed prediction of both the best vaccine match for any single virus and the breadth of coverage of new vaccine candidates from their capsid sequences as effectively as or better than serology. Sub-sequences chosen by the model-building process all contained sites that are known epitopes on other serotypes. Furthermore, for the SAT1 serotype, for which epitopes have never previously been identified, we provide strong evidence – by controlling for phylogenetic structure – for the presence of three epitopes across a panel of viruses and quantify the relative significance of some individual residues in determining cross-neutralization. Identifying and quantifying the importance of sites that predict viral strain cross-reactivity not just for single viruses but across entire serotypes can help in the design of vaccines with better targeting and broader coverage. These techniques can be generalized to any infectious agents where cross-reactivity assays have been carried out. As the parameterization uses pre-existing datasets, this approach quickly and cheaply increases both our understanding of antigenic relationships and our power to control disease. New strains of viruses arise continually. Consequently, predicting when past exposure to closely related strains will protect against infection by novel strains is central to understanding the dynamics of a broad range of the world's most important infectious diseases. While previous research has developed valuable tools for describing the observed antigenic landscapes, our ability to predict cross-protection between different viral strains depends almost entirely on cumbersome and expensive live animal work, often restricted to model species rather than the natural host. The development of computer-based approaches to the estimation of cross-protection from viral sequence data would be hugely valuable, and our study represents a significant step towards this research goal. The genetically highly variable nature of RNA viruses in vitro and subsequently in vivo. Due to the time and expense required, there is a limit to the number of isolates that can be submitted to undergo this procedure, and a sub-optimal choice of vaccine strain may therefore be made. An in silico predictor that identifies those strains likely to provide the broadest cross-protection could therefore substantially enhance capacity to develop appropriate vaccines rapidly and effectively, whilst minimising the cost and the need for animal experimentation.The emergence of antigenically novel viruses, against which existing vaccines do not provide adequate protection, may require the selection of new vaccine seed strains. Currently, where no appropriate vaccine exists, field isolates are, when possible, adapted for vaccine production, amplified and then processed into vaccines Apthovirus of the family Picornaviridae. It exhibits great genomic variability, with 32–33% and 53% amino acid variability in our data, within and between serotypes respectively, in the immunogenically important structural proteins, VP1-VP3 . Amongst the exposed residues are epitopes recognised by the host immune system. All serotypes are believed to share the major antigenic site on the flexible G-H loop of the VP1 protein, which is highly variable between even closely related strains. This is the only site to have been identified with monoclonal antibody (MAb) escape mutants for a single SAT2 virus in vitro measure of whether the sites that contribute to the neutralization of the virus remain sufficiently similar to cross-react. Virus neutralisation is not the only important determinant of protection 1-values” – the ratio of the heterologous to homologous titres, with a ratio close to 1 indicating the viruses are antigenically similar. Generally r1-values in the range of 0.4–1.0 are considered to be indicative of reasonable levels of cross-protection, whilst all values being below 0.2 for a given isolate indicate the need for new vaccine strain development 1-values do not fully account for these different sources of variability This antigenic variability is reflected in the virus neutralisation (VN) titres The availability of sequence data and related titres from VN testing provides the opportunity to directly relate cross-reactivity to sequence variation. This relationship would allow prediction of an important component of vaccine efficacy for candidate vaccine seed strains, and rapid identification of vaccine match without the need for new serology work for existing vaccines.in silico tool to predict vaccine efficacy using sequence data, neutralising titres and structural information, and use this tool to identify and quantify the significance of epitopes of the viruses. We have obtained a broad spectrum of SAT1 and SAT2 isolates which were sequenced, and have generated sera from representative viruses. An extensive serological dataset was generated from VNTs. We have also used a novel, and currently the only, crystallographic structure for any SAT serotype to identify surface-exposed residues on the capsid. Specific objectives were to (i) generate improved statistical methods of estimating r1-values that maximise the efficient use of available experimental data, (ii) relate these estimated antigenic differences to sequence variation, (iii) use this relationship to predict vaccine match for viruses from sequence information, and to predict neutralisation titres, cross-reactivity and hence coverage for vaccine strains, and then (iv) to identify areas of the capsid containing epitopes.The aim of the current study is to develop an Twenty SAT1 and twenty-two SAT2 viruses  represen1-value controls for between-serum variability 1-values simultaneously.A key feature of our analysis was to develop a formal approach for including data from multiple sera, experiments and batches. The use of homologous and heterologous titres from the same serum to calculate an r−44), protective strain (p = 10−12) and their interaction (p = 10−27) were significant fixed effects, but neither serotype nor whether sera were prepared by vaccination or infection was found to be significant. We would not have expected to see this latter effect since it was confounded with protective strain as all sera for each strain were generated by only one of vaccination or infection. A random effect at the level of experiment accounted for the inter-experiment variability. By comparing models with random effects to allow for other sources of variability, we determined that there was consistent variability between sera (p = 10−15), but not between batches  variances sum to 0.287, which gives a 95% confidence interval around the estimate of an 1-valueindividual serological r  of +/−1.49 log titres. With test variability this high it is clear that an improved method of estimating r1-values that makes use of all available data would be valuable.Estimates of these variances were used to examine the expected uncertainties associated with standard methods of estimating rp,cμ had 95% confidence intervals ranging between +/−0.50 to +/−1.17 log titres, depending on the number of titres available . These narrower confidence intervals show that our new techniques for estimating p,cμ provide a substantial improvement on existing methods for the same number of titres. These best consensus estimates of the true means were therefore used as our gold standard for subsequent analyses  – was replaced with a predictive term based on differences between capsid-coding sequences of the protective and challenge strains. Removing this fixed effect also necessitated the inclusion of the additional random effect of challenge virus (−10), and the final model took the form:k0 is the average titre and di is a raw count of the number of amino acid changes between the protective strain and challenge virus in a single candidate area identified from the structural modelling  the number of times the protective strain with the predicted highest r1-value for a given challenge virus matches the strain that would be selected using the serology data; and (ii) the difference between the predicted r1-values for specific pairs of protective strains and challenge viruses and our best consensus estimate r1-values for those pairs. In the former case only those challenge viruses that might have appropriate protective strains are considered (we chose those with an estimated r1≥0.2) since we are not interested in whether the model correctly chooses the least worst vaccine when none could possibly be effective.Two different predictions are of interest: first, in an outbreak, the vaccine that best matches a given challenge virus, and second, to judge breadth of coverage of a candidate vaccine strain – the range of r1-values generated for that missing protective strain compared with our gold standard estimates. The best models for each serotype are reported below.Specifically, candidate models selected by the model-building process were cross-validated by estimating parameters in two different ways, corresponding to our two different requirements: (i) using datasets missing all data for each challenge virus in turn, and comparing the vaccine choice with that obtained using our gold standard estimates; or (ii) using datasets missing all data for each protective strain in turn, and comparing the r1≥0.2), and so were included in the cross-validation. For SAT1 the best model after cross-validation contained two terms, the number of amino acid changes in the VP1 G-H loop and beyond , and in the VP3 H-I loop . Though additional serological and sequence data would ultimately improve the predictive model, it currently performs at least as well as standard serological approaches.The predictive models successfully identified 13 SAT1 matches (72%) and all 9 SAT2 matches (100%) . This ac1-values for different protective strains using the same challenge virus) potentially at the expense of absolute error . In this case, however, for accurate r1-value prediction the aim is to reduce this absolute error. Adding a term to the models that explicitly distinguished homologous and heterologous titres removed this bias, and so it was included in all of the candidate models.A small bias was observed for all of the candidate models in their predictions for heterologous titres relative to homologous titres. This does not affect the vaccine match experiments where the aim is to reduce relative error , together with a term that is present when titres are heterologous. The formula is:The best predictive model of r1-values (grey dots) at 87% . The variance around the gold standard estimates is significantly lower for the predictions than the individual serological r1-values .Ninety eight percent of the predictions , again together with a term for heterologous titres. The formula is:The best predictor of r1-values were within the confidence limits of the gold standard estimates . Variances were also not significantly different (0.40 compared to than 0.43). Both predictions and serological measurements are less accurate than those for SAT1 due to the lower repeatability of SAT2 serology . Each branch partitions the tree into two groups, and where a branch effect represents changes that impact significantly on cross-reactivities, they will be higher between viruses within the groups than those between groups (after controlling for other effects). For instance, where a terminal branch is identified, the fixed effect of that branch specifies an amount by which the virus to which it leads (the first group) is antigenically distant from all other viruses (the second group). A significant internal branch, similarly, identifies a clade that is antigenically distant from the rest of the tree. By building a model containing all of the branches in the tree, and then using a stepwise elimination procedure to remove branches which do not significantly improve the model fit (p>0.05), we are left with the set of branches that, when traversed, significantly account for reductions in antigenic cross-reactivity.The above predictive models identify those areas that are correlated with loss of cross-reactivity. To identify those areas that are directly responsible for antigenic variability it is necessary to develop models that additionally control for the phylogenetic relationships between virus strains. The phylogenetic control extends Equation 1 in an analogous manner to the predictive model (Equation 2):1-value above 0.2). For SAT2, there are six internal branches throughout the tree and fifteen terminal branches leading to ten of the thirteen viruses that are antigenically distinct  and five other viruses (including the four protective strains).Twelve phylogenetic branches are significant in SAT1 and twenty-one in SAT2 . Of these eight terms, seven were individually significant in the previous test (p<0.05). These seven terms consisted of five that corresponded exactly to the five areas identified as the constituent parts of the Site 3 conformational epitope for A10 −8 collectively), one was the VP1 G-H loop (p<0.001), and the last was the VP3 G-H loop, previously identified as Site 3 on A12 Because substitutions are ultimately responsible for the loss of cross-reactivity, the substitutions contributing to counts in these SAT1 areas must be responsible for this loss unless they are co-occurring with causative substitutions. Any such causative substitutions should, however, be identifiable because substitution counts in areas containing them will improve the model fit. Comparing the individually best SAT1 areas from above with bootstrapped random sequences of the same length from other parts of the capsid, however, fails to identify other causative substitutions, and eight areas were instead found to be significant after a Holm-Bonferroni correction for the number of terms To identify the specific residues responsible for these drops in antigenic cross-reactivity, Equation 8 is trivially modified to test substitutions to the 62 individually variable residues in the seven areas identified (instead of the 29 candidate areas). Again, this is a conservative test, as it will only identify residues where multiple/convergent substitutions occur at different branches in the phylogeny. Collectively, changes to the residues are significant after controlling for phylogeny (p<0.005), but only two residues are individually significant (p<0.05). Bootstrap comparisons with other residues showed these to be the two most significant predictors of loss of cross-reactivity out of all the residues in the capsid, and both are adjacent to residues identified by MAb escape mutant studies on A10 as part of Site 3 κ(r) in Equation 9) that allows the count (k1 in Equation 8) to vary in significance for different protective strains, challenge viruses or sera (r) can be added to the model:To test whether areas and individual residues vary in their significance across the whole serotype or are conserved, a random effect . For SAT2, however, although no areas or residues are significant individually, thirty-one of them have significant interactions (p<0.05) with at least one of these ‘The identification of antigenic sites on individual FMDV isolates is time consuming, with the consequence that data are not available for all serotypes, much less for all isolates. Indeed, very little is known about the important epitopes for SAT1 and SAT2 viruses, impacting on the potential to both design vaccines with broader or better targeted antigenic cover and predict the efficacy of a particular vaccine strain against circulating viruses in the field. We have identified seven areas containing what we believe to be three epitopes for SAT1, and we provide evidence that these are conserved across our whole sample. We have further quantified the effect of substitutions at two specific residues in one of these epitopes. The conservative phylogenetic control employed throughout the analysis means that this may not be an exhaustive list of antigenically significant areas of the capsid, and is almost certainly not for residues, as it will only identify ones where multiple/convergent substitutions occur at different branches in the phylogeny. The areas that are identified do, however, correspond to epitopes identified by MAb escape mutants for other serotypes, and both of the specific residues found are  adjacent to ones which are part of Site 3 on A10 We have also used cross-reactivity data generated for SAT1 and SAT2 FMDV to develop a linear mixed-effects model that uses replicates and repeated experiments to more accurately account for variability in measurement and so generate better estimates of cross-reactivity for FMDV. We have enhanced this model with sequence and structural data to identify surface-exposed residues that correlate with loss of cross-reactivity and then built models using counts of amino acid substitutions in selected areas to predict cross-reactivity. We note that all of the areas used in these models are associated with epitopes identified by MAb escape mutants for other serotypes. Furthermore, for SAT1 they also correspond to parts of the new epitopes identified above.1-values, and 77% of SAT2 predictions. For SAT1 this was significantly better than individual serological r1-values despite the predictions being made without the use of any sera from the protective strains. Related models were also used to predict the best vaccine match for novel virus isolates. For the 9 SAT2 virus isolates for which any match existed in the data (r1≥0.2), the model correctly predicted all 9, and for the 18 SAT1 isolates, the model predicted 13. In both cases there is no significant difference between the model predictions and the serological results. The uncertainty inherent in the VNT and the variability between different experiments balance any inaccuracies in the predictive model, making it at least as effective a measure of r1-value and vaccine match as serology itself unless the latter is repeated multiple times. Improving the serological tests is an area of active research These predictive models were used to successfully identify the efficacy of novel candidate vaccines against the virus isolates, with 98% of SAT1 predictions within the 95% confidence intervals for our gold standard estimator of true r1-value predictions and the inaccuracy of the matching for SAT1 (relative to SAT2) may have the same cause – two of the anti-sera were raised against viruses of the same topotype (topotype 1 – constituting 70% of the titres), producing a good model for that topotype, but with little power to generalise and predict cross-reactivity for significantly different viruses clustering in other topotypes. Consequently, it identified 8 out of 10 correctly when the answer was a topotype 1 vaccine, but only 5 out of 8 for the other topotype. Conversely, the relative inaccuracy of the r1-values but the accuracy of the matching for SAT2 may share the opposite cause – 4 anti-sera were raised against 4 different topotypes, giving a better estimate of which areas were antigenically significant in general and thereby allowing better vaccine matching. However, because of the relative sparsity of data from any individual topotype , we obtained a poorer estimate of the relative importance of each area, and therefore a less accurate r1-value prediction. The greater inherent variability in SAT2 titres also necessitated more data to acquire an accurate estimate. These complementary results suggest that to refine the model further more data from different topotypes will improve the vaccine matching in SAT1, and more from the same topotypes will improve the estimates of cross-reactivity in SAT2.The accuracy of the rThe VP1 G-H loop is known to contain major epitopes in all serotypes of FMDV where MAb studies have been conducted; substitutions in it are therefore considered to be a significant determinant of loss of cross-reactivity. Our SAT1 epitope analysis identified this loop and the model used it to predict cross-reactivity. However, our SAT2 model did not find the number of amino acid substitutions in the candidate area containing the loop to be in general a good correlate. There are three potential explanations. First, the G-H loop has a high substitution rate, so high indeed that we suggest that the epitope(s) may very often not cross-react even between closely related strains, giving the model little data with cross-reacting epitopes from which to identify a pattern. Second, the candidate area is much bigger than just the G-H loop, therefore including many residues that are not antigenically significant. Finally saturation can occur, causing the actual count even inside the G-H loop to cease to be meaningful. The combination of these effects makes it very difficult to detect the signal of epitope loss from the noise of other substitutions.a priori criterion, the strategy for choosing models to test was fixed in advance, and the relatively small number of models examined in the model-building process was easily controlled for by a simple statistical correction.Better predictors might be obtained by selecting smaller candidate areas. However, there is a risk of “fishing” until an appropriate sub-sequence is discovered, leaving the generality of the technique uncertain. Related work on influenza A A further strength of our approach is that it identifies antigenically significant areas. These can be validated entirely independently through comparison with MAb escape mutant-derived epitope information. Of the 5 areas identified here, all have been identified by previous MAb work in other serotypes. Furthermore, the two SAT1 areas are specifically identified in separate analyses as containing epitopes for this serotype, which provides further strong evidence that the model is indeed predictive rather than merely correlative. For SAT2 the evidence is weaker since the epitopes have only been identified on other serotypes, and the evidence that there may be epitopic variability within the serotype suggests that more caution should be applied in using them to predict the cross-reactivity of distantly related isolates.This work could be further validated by a reverse-genetics approach targeting the specific SAT1 residues identified as antigenically significant. Our methodology could be used to evaluate the effect of amino acid substitutions by predicting coverage against a panel of circulating viruses, allowing potential vaccine candidates that are expected to better match the panel to be easily identified. Interestingly, because of the high substitution rate obscuring any signal in the VP1 G-H loop, identified substitutions would probably not be located in the major antigenic site of FMDV but instead be found in other antigenically important areas.The technique developed here can be used directly for any FMDV serotype and potentially for any similar virus where cross-reactivity, sequencing and structural studies have been carried out, both to identify epitopes, and to predict vaccine match for new isolates and estimate efficacy of new candidate seed strains. This can be done by exploiting historical datasets, and is therefore a quick, low cost and valuable method for better understanding antigenic relationships. In summary, the use of sequence data to predict antigenic relationships is a powerful tool that has the potential to be applied to a variety of different infectious agents.All procedures were approved by the Onderstepoort Veterinary Institute Animal Ethics Committee according to national animal welfare standards.ca. 2.2kb region were compiled using Sequencher™ vs4.7 (Gene Codes Corporation). Since these are protein-coding regions, the amino-acid sequences were aligned with ClustalW (v.1.83) and this alignment was then used to align the nucleic-acid sequences.The viruses were either supplied by the World Reference Laboratory for FMD at the Institute for Animal Health, Pirbright (United Kingdom) or form part of the virus databank at the Transboundary Animal Diseases Programme (TADP), Onderstepoort Veterinary Institute (South Africa). Viral RNA was extracted from cell-culture-adapted isolates and cDNA synthesised 50 in 50% of the wells(ibid.).The antigenic diversity of the field isolates was determined using virus neutralisation assays in micro-titre plates using IB-RS-2 cells as the indicator system The crystal structure of the SAT1 BOT/1/68 capsid was solved at a resolution of 3Å  . This is the only structure in existence for any SAT1 or SAT2 virus. SAT1 amino-acid sequence alignments were compiled with ClustalW. Structures were visualised and the surface-exposed residues identified with the PyMol Molecular Graphics System v1.2r0 (DeLano Scientific LLC). Exposed regions of SAT2 were approximated by alignment with the SAT1 structure.candidate areas each of which might be implicated in one or more antigenic site(s). Of these, 14 of the sub-sequences were invariant in SAT1, as were 15 of the areas in SAT2, leaving 29 and 28 areas respectively in the two serotypes to test as possible predictors  or infection (0);Serum: each serum is drawn from a single animal, so serum labels correspond to individual donor animals;Challenge virus: virus isolate used in neutralisation experiment;Serotype: serotype of the protective  strain;experiments where the same serum was used at the same time with the same challenge virus (tests within an experiment are replicates); different experiments were grouped into batches by the time at which they were performed. Possible variability at these levels was investigated when building the model. Our gold standard best consensus estimates of r1-values are available in The neutralisation tests were grouped into There were five stages to the statistical modelling:−15) nor homoscedastic residuals 1-values from serological data.First, a linear mixed-effects model The fixed effects in this model (Equation 1) were then replaced with sequence-based predictors , and the model selection approach outlined below used to generate a model that predicted cross-reactivity directly from sequence data (Equations 2–6).These were replaced in turn with phylogeny-based effects (see below) to control for the phylogenetic structure of the data (Equation 7).Individual areas and residues were added to this model to identify epitopes (Equation 8).Finally, within-serotype variability in epitopes was investigated (Equation 9).i, the statistic −2 Σ log pi is expected to be χ2 distributed with twice as many degrees of freedom as tests under the null hypotheses The predictive models (Equation 2) were generated by sequentially adding the count of non-synonymous changes in each of the candidate antigenic areas as a fixed effect in a standard stepwise regression which continued for a variable number of steps until no further terms could be added. Specifically, the probability that each model was a significantly better predictor than its precursor was assessed by a likelihood ratio test since the models were nested. For multiple tests each with p-values pAmino acid substitutions on the capsid (including those identified above) are correlated with antigenic distance; this could be a direct relationship or may arise indirectly via relationships between substitutions, phylogenetic history and antigenic drift, as is found in influenza A contrasts) that are used to control for the evolutionary history, whereas for us these are the signal – the cross-reactivity. We wish instead to identify the causes of the changes in cross-reactivity while controlling for the common evolutionary history.However, existing mechanisms for controlling for phylogeny focus on properties (or traits) of the leaves of tree and not the relationships between them δi, Equation 7); this is non-zero if the branch is travelled (and thus these changes have occurred) in the traversal of the tree between the protective strain from which the serum is derived and the virus isolate in a cross-reactivity test. Including these terms in the analysis controls for repeated measures of this traversal.In practice, each branch on the phylogenetic tree (see above) represents a set of common substitutions by which any pair of viruses either side of the branch differ (unless multiple and/or convergent substitutions have occurred). Any comparison of antigenicity between two viruses either side of the branch will be affected by those changes. A fixed effect is therefore added to the model for each branch . These ambiguities mean that the models cannot be used predictively, but this does not prevent their use for phylogenetic control.Model development begins by constructing a model with the maximal set of fixed effects; these are then removed using stepwise regression until all of those left significantly improve the model fit (p<0.05). In this manner we have controlled for repeated measures of every significant piece of shared phylogenetic history. Because the phylogenetic trees are different for the two serotypes the serotypes are modelled separately.The phylogenetic control terms account for repeated measurement of all significant shared phylogenetic history. However, in doing so they remove all significant direct effects of substitutions at individual branches of the tree. Consequently, the substitution count in any area can only significantly improve the model if it corresponds to multiple and/or convergent substitutions at the same sites in different branches. Modelling phylogenetic control in this way therefore provides a conservative estimate of the number of areas that directly affect antigenicity.d1, Equation 8), and these were then compared to bootstrapped samples from the remaining capsid surface (randomly assembled sub-sequences of the same length as each candidate). Since changes to the capsid proteins must be responsible for loss of cross-reactivity, comparing specified sub-sequences to the rest of the capsid after controlling for phylogenetic structure directly determines whether these areas contain true predictors or whether they contain correlates, with the true epitopes being found elsewhere. To identify the individual constituent residues of epitopes, exactly the same mechanism is used on individual residues instead of areas.Sequence-based predictors were added to the models – again, substitution counts for each of the candidate areas identified in the structural analysis – to determine which sub-sequences were the best predictors after controlling for phylogeny . Should one of these terms be found to improve model fit, this would provide evidence that there is variability across the phylogeny as to the importance of different sites.Dataset S11-values used in the study.Best consensus estimates of all r(0.03 MB XLS)Click here for additional data file.Dataset S2Reference alignment of the protective strains to the study isolates. The dataset shows the VP2, VP3 and VP1 proteins of the protective strains after alignment to all of the SAT1 and SAT2 isolates used in the study. The 43 contiguous surface-exposed areas identified by the capsid structural analysis are highlighted. Areas with no amino-acid variability are in grey, and areas with variability are in red for the serotypes for which there is variation.(0.14 MB DOC)Click here for additional data file.


Latino children are at increased risk for mirconutrient deficiencies and problems of overweight and obesity. Exposures in pregnancy and early postpartum may impact future growth trajectories.To evaluate the relationship between prenatal and postnatal maternal depressive symptoms experienced in pregnancy and infant growth from birth to 2 years of age in a cohort of Latino infants.We recruited pregnant Latina mothers at two San Francisco hospitals and followed their healthy infants to 24 months of age. At 6, 12 and 24 months of age, infants were weighed and measured. Maternal depressive symptoms were assessed prenatally and at 4-6 weeks postpartum. Women who had high depressive symptoms at both time periods were defined as having chronic depression. Logistic mixed models were applied to compare growth curves and risk for overweight and underweight based on exposure to maternal depression.We followed 181 infants to 24 months. At 12 and 24 months, respectively, 27.4% and 40.5% were overweight, and 5.6% and 2.2% were underweight. Exposure to chronic maternal depression was associated with underweight  and with reduced weight gain in the first 2 years of life  compared with unexposed infants or infants exposed to episodic depression (depression at one time point). Exposure to chronic depression was also associated with reduced risk for overweight in the first 2 years of life .Exposure to chronic maternal depression in the pre- and postnatal period was associated with reduced weight gain in the first two years of life and greater risk for failure to thrive, in comparison with unexposed infants or those exposed episodically. The infants of mothers with chronic depression may need additional nutritional monitoring and intervention. Maternal prenatal depression can have significant and critical negative impact in fetal development and occurs in approximately 7 to 13% of childbearing women In addition to the above described associations between prenatal and postnatal depression, lower birthweight, and failure to thrive, maternal postnatal depressive symptoms have also been associated with risk for pediatric overweight in the first 2 years of life. In a cross-sectional study from Brazil, Surkan et al, reported that postnatal depressive symptoms (from 6-24 months) are associated with increased risk for overweight and obesity for children between 6 and 24 months of age Studies suggest that the time period of exposure is likely important in determining childhood outcome. A recent US-based study found that children exposed to antenatal maternal depression have a lower BMI z-score at 3 years of age, while those exposed to postnatal maternal depression at 6 months have higher overall adiposity No study to date has evaluated the combined impact of exposure to prenatal and postnatal depression in a population at high risk for future obesity such as Latino infants, in comparison with unexposed infants or those exposed to maternal depressive symptoms either prenatally or postnatally. Such an approach is needed as the infants of women who have more intractable depressive symptoms that span the perinatal period may have different growth trajectories in comparison with infants who do not have these environmental exposures or a more limited form of exposure .th percentile), at 5.7%, compared with 3.1% of non-Hispanic whites The early growth trajectories of Latino children are an important group to assess given the high prevalence of nutritional deficiencies in this population as well as the high risk of future obesity. Latino children have high prevalences of zinc deficiency nd and 3rd trimesters of pregnancy at prenatal clinics at the University of California, San Francisco (UCSF) Medical Center and San Francisco General Hospital (SFGH) from May 2006 to May 2007, and were followed through labor and delivery until the infants reached 2 year of age. Exclusion criteria included drugs or alcohol abuse, pre-existing diabetes mellitus or gestational diabetes mellitus treated with insulin, polycystic ovarian syndrome, any eating disorders such as bulimia or anorexia nervosa, or any health problems that would influence breast-feeding. Infants at delivery were excluded if they had special care needs, chronic disease, or Apgar scores of less than 7 at 5 minutes. Close to 100% of Latina women were approached in both hospitals as we had a continuous on-site research presence. Approximately 90% of those who met the inclusion criteria for the study agreed to participate.Latina women were recruited during the 2All procedures were approved by the Committee on Human Research at UCSF and the Institutional Review Board at SFGH. Following informed consent, baseline data and socio-demographics of the participants including age, education, occupation, income, marital status, language use and length of time in the United States were collected. Medical history was also ascertained through chart review and by questionnaire to determine mental health history, including previous diagnosis and/or treatment for depression or anxiety. Maternal pre-pregnancy weight was collected at baseline by self-report. Questions on legal status or migration history were not asked other than length of time in the United States because of the sensitive nature of these questions and the possibility that questioning in this area could compromise the validity of our study or jeopardize the follow-up rate. Additional specifics about our study population are described in Wojcicki et al. Upon enrollment, mothers were administered the Edinburgh Postpartum Depression Scale (EPDS) 2). Waist circumference in children was measured at 24 months of age in a sub-sample of the cohort (n = 156). A small percentage of the cohort (between 3–5% depending on timepoint) were not available to be measured so weight and height were extracted from the medical record. BMI <25 kg/m2 was categorized as normal/underweight, 25≤ BMI <30 was categorized as overweight, and BMI ≥30 was categorized as obese, using standard definitions from the World Health Organization At birth, anthropometric measurements of the infant, including infant weight  and length (using standard tape measurements) were obtained. Gestational age and Apgar scores were recorded. At 4–6 weeks postpartum, participants were contacted by phone and interviewed again for depressive symptoms and clinical depression using the same instruments employed at baseline. Additionally, infant feeding was assessed using a 24 hour dietary recall. At 6, 12 and 24 months post-partum, the infants were weighed and measured. Maternal weight  and height (using a portable stadiometer to nearest 0.1cm) were measured between 12 and 24 months postpartum to calculate body mass index (BMI) (kg/mth percentile weight-for-age), overweight (≥85th percentile weight-for-length or body mass index) and obese (≥95th percentile weight/height or body mass index) as described in more detail below. Additional predictor variables considered for multivariate regression models included infant birth weight z-score or birth weight-for-length z-score, any breast-feeding at 6 months of age, maternal postnatal (12–24 months) BMI, maternal ethnicity , maternal age and gestational age. Additional details about specific analyses are described below. Birthweight andg birthweight for length Z scores were adjusted for in the analyses because our primary outcome of interest was weight gain from 6 months to 2 years of age in relation to prenatal and postnatal maternal depression exposures.The main predictor of interest was depressive symptoms experienced either prenatally and/or at 4–6 weeks postpartum described more in detail below. The main outcome was child weight-for-length z-score at 6,12, and 24 months of age. Actual ages at post-partum visits varied somewhat from these values, so analyses used ages from observed visits. Analyses based on the grouped ages  yielded very similar results (not shown). Secondary outcomes of interest included risk for underweight/failure to thrive  CESD ≥16; 2) EPDS ≥13; or 3) having a major depressive episode or dysthymia as per the MINI. A high depression symptom score was defined as a high score on one of the above measures prenatally or at the 4–6 week timepoint. Chronic depression was defined as having a high score at both time points  and episodic depression was defined as having depression at one time point . We also evaluated the relationship between weight trajectories and not ever having depressive symptoms compared with having depressive symptoms only in the prenatal period, only in the postnatal period, and at both time-points.th percentile BMI or weight/length) and obesity (≥95th percentile BMI or weight/length) up until 24 months were defined by the National Center for Health Statistics, Center for Disease Control and Prevention weight-for-length and body mass index growth curves th percentile weight) was defined using the same growth curves. Z-scores were similarly defined from the National Center for Health Statistics, Center for Disease Control and Prevention th percentile) was defined using the cut-points described by Fernandez et al. Overweight  and Student's t-tests were applied to evaluate the relationships between exposure to prenatal and/or postnatal depression and weight/length percentiles, risk for underweight and risk for overweight and obesity. Mixed effects linear regression models were used to evaluate differences in weight-for-length z-scores over the two-year period in relation exposure to maternal depression adjusting for the potential confounders described above. Models included random intercepts, and random slopes for time effects. Differences in individual weight trajectories between groups reporting different levels of depression were evaluated via inclusion of appropriate interaction terms in fitted models. Mixed effects logistic regression models were used to determine risk factors for failure to thrive, overweight and obesity. Time effects were modeled using the same approach described for linear models. Outcomes for logistic and linear models only included data from visits occurring post-birth as our major predictors included postnatal depression at 4-6 weeks postpartum. As birthweight occurred prior to postnatal depression, it was not possible to include birthweight in the longitudinal analysis as an outcome because of the temporal sequence of events.Only women who had all data on the selected variables were included in the multivariate models. Data were entered into Excel, and subsequent analyses were conducted using Stata 11.0 .ice. Five imputed datasets were created and the command mim was used to run mixed logistic and mixed regression models with complete datasets similar to the models described above. Approximately 8% of the data was missing from the bivariate models and 15% from the multivariate models. The bivariate and multivariate results were similar for our primary outcomes comparing the unimputed and imputed models.In order to assess the possible role of missing data on our estimates, multiple imputation methods were used in Stata 9.0 using the command 18 and followed 175 women to 6 months of age and 174 to 12 months of age and 181 to 2 years of age. The majority of our cohort (61.2% [123/201]) was Mexican ethnicity; the remainder was primarily of Central American origin . Most women (92.0%) were participants in the Women, Infants and Children's Program (WIC), 31.0% were married, and 52.0% were living with partner. Our participants tended to have a high school diploma or less (76.0%), and 92.9% cited Spanish as their primarily language. The mean number of other children in the household was 0.83±0.99. The majority of the infants in the cohort were born vaginally (85.6%). Mean gestational age was 39.3 weeks, and mean birthweight was 3.37±0.48 kilograms as described in Wojcicki et al. We enrolled 201 women in the cohortPrenatally, 28.9% of participants had depressive symptoms or clinical depression. At 4–6 weeks post-partum this number declined to 15.7%. Depressive symptoms were evident in 22.5% of women at the prenatal or 4–6 week time point, and 11.0% had chronic depression at the prenatal and 4–6 week timepoint. A smaller 4.0% (8/201) had clinical depression prenatally and 4.3% at 4–6 weeks but only one of these eight had clinical depression at both time-points.th percentile for weight. By 2 years of age, only 2.2% were <5th percentile for weight  of the mothers were breast-feeding (BF) at 4–6 weeks, 70.0% at 6 months, and 38.8% at 12 months of age . A much th percentile at age 2 years (results not shown). Exposure to chronic maternal depression was associated with decreased frequency of overweight at 6, 12 and 24 months, although the results did not achieve statistical significance . SimilarLogistic mixed modeling revealed that exposure to chronic depression within the first two years of life was associated with a decreased weight-for-length z-score  compared with no exposure or exposure only episodically . In modeSimilarly, exposure to chronic maternal depression was associated with increased risk for failure to thrive/underweight  . In multWhen exposure to pre- and post-natal depression was evaluated separately in relation to weight-for-length z-scores and risk for failure to thrive, we did not find differences between the models that evaluated exposures to prenatal and postnatal depression separately and those that evaluated episodic versus chronic exposure to maternal depression. In both models, increased risks for lower weight-for-length z-scores and failure to thrive were associated with exposure to chronic depression compared with infants who were unexposed, exposed to episodic depression, or exposed to prenatal or postnatal depression alone. We also did not find any effect using an interaction term modeling time in relation to maternal depression status so we removed the interaction term from the multivariate models.Ours is the first study to longitudinally evaluate the relationship between weight gain patterns in the first two years of life in Latino children and exposures to maternal depression in the perinatal period. In contrast to other studies that have evaluated these relationships Other studies have evaluated relationships between depression and growth either cross-sectionally th week of life until 9 months of age compared with unexposed infants Our results also revealed that infants exposed to prenatal depression in utero or to chronic depression were more likely to have failure to thrive in the first 2 years of life compared with children who were unexposed. Previous studies have found that high prenatal cortisol is associated with prematurity and lower birthweight through hyperactivation of the HPA axis While we did not find any differences in exposure to depression and breastfeeding duration, previous investigators have reported that women with depressive symptoms in the postpartum period have decreased breastfeeding duration and increased breast-feeding difficulties We did not investigate the relationship between medication use for treatment of depression on infant outcomes because of the low percentage of medication use in our population, although this could be a fruitful area for future research. Similarly, we also did not examine or the role of stress hormones such as cortisol or growth hormones on patterns of weight gain.From a clinical standpoint, infants of women with chronic depression should be evaluated for failure to thrive and the mothers should potentially be provided with additional nutritional support in addition to any needed psychological or psychiatric intervention. Interventions should be proactive given the behavioral and developmental risk factors associated with failure to thrive. Our results indicate that it is the combined exposures in utero and maternal-child interactions in the postpartum period will also contribute to failure to thrive, in contrast to other studies which have only assessed the role of prenatal depression or postnatal on child weight gain


To investigate the multiple effects of NO3- pollution on microbial communities, we created two shotgun metagenomes from vernal pool microcosms that were either enriched with a solution of 10 mg NO3--N (+NO3-) or received distilled water as a control (−N).Human activities have greatly increased nitrogen (N) levels in natural habitats through atmospheric N deposition and nutrient leaching, which can have large effects on N cycling and other ecosystem processes. Because of the significant role microorganisms play in N cycling, high inputs of nitrogenous compounds, such as nitrate . Surprisingly, we found no changes to N metabolism EGTs, even though large shifts in denitrification rates were seen between the +NO3- and –N microcosms. Thus, in the absence of NO3- addition, it is plausible that the microbes used other respiratory pathways for energy. Respiratory pathways involving iron may have been particularly important in our –N microcosms, since iron acquisition EGTs were proportionally higher in the –N metagenome. Additionally, we noted a proportional increase in Acidobacteria and Alphaproteobacteria EGTs in response to NO3- addition. These community shifts in were not evident with TRFLP, suggesting that metagenomic analyses may detect fine-scale changes not possible with community profiling techniques.After only 20 hours of exposure to NO3- pollution on microbial communities, and the relationship between community stability and function in response to disturbance.Our results suggest that the vernal pool microbial communities profiled here may rely on their metabolic plasticity for growth and survival when certain resources are limiting. The creation of these metagenomes also highlights how little is known about the effects of NO A BLASTX comparison to a non-redundant protein database was used to match the EGTs in the metagenomes to SEED subsystems [-5 or lower and had sequence similarity of 50 base pairs or greater were included in our MG-RAST analysis.Because the metagenomes constructed from our microcosms contained DNA reads from multiple species, they were analyzed unassembled using the MG-RAST server  and are bsystems . The SEEbsystems ,21,51 anbsystems ,21,49-51amoA, amoB, and amoC) and nitrogen fixation  were also included. The database contained a total of 111,502 sequences and a complete list of the genes included in the database can be found in Additional file nosZ”.) The local BLASTN was conducted at Case Western Reserve University’s Genome and Transcriptome Analysis Core facility. A number of sequences in our database were complete chromosome sequences that included genes other than the N metabolism genes we were interested in. If sequences from the metagenomes matched with these database entries, they were only retained if the gene region of the BLASTN match was to a N metabolism gene of interest  The BLASTN comparison included an e-value cutoff of 10-5 or lower and sequence similarity cutoff of 50 base pairs or greater.Metagenomes were also analyzed with a local BLASTN to a database of N metabolism genes that we constructed with searches at the NCBI site. The database included the known genes for the enzymes involved in denitrification, DNRA, and Annamox  program was used to compare the +NOct tests . The MG-ct tests . ConfideThe authors declare that they have no competing interests.SC-K conceived of the study, collected and processed samples for sequencing, and authored the manuscript. KS participated in the design and implementation of the study and edited and commented on the paper. DB conceived of the study and participated in its design and implementation, contributed to data analysis, and edited and commented on the paper. All authors read and approved the final manuscript.Results from Fisher exact tests at all subsystem levels and a chi-square test conducted at level two using the Statistical Analysis of Metagenomic Profiles program.Click here for file3- metagenome to nitrogen metabolism genes with a BLASTN.Nitrogen metabolism genes included in the database created from the NCBI site and all matches from the +NOClick here for file