A multivariate random-effects meta-analysis must incorporate and estimate the between-study correlation . We consider two applied examples, one involving a diagnostic marker and the other a surrogate outcome. These motivate a simulation study where estimation properties from BRMA are compared with those from two separate univariate random-effects meta-analyses (URMAs), the traditional approach.ρB as -1 in both applied examples. Analytically we show this is due to the maximum likelihood estimator sensibly truncating the between-study covariance matrix on the boundary of its parameter space. Our simulations reveal this commonly occurs when the number of studies is small or the within-study variation is relatively large; it also causes upwardly biased between-study variance estimates, which are inflated to compensate for the restriction on B. Importantly, this does not induce any systematic bias in the pooled estimates and produces conservative standard errors and mean-square errors. Furthermore, the normal BRMA is preferable to two normal URMAs; the mean-square error and standard error of pooled estimates is generally smaller in the BRMA, especially given data missing at random. For meta-analysis of proportions we then show that a generalised BRMA model is better still. This correctly uses a binomial rather than normal distribution, and produces better estimates than the normal BRMA and also two generalised URMAs; however the model may sometimes not converge due to difficulties estimating ρB.The normal BRMA model estimates A BRMA model offers numerous advantages over separate univariate synthesises; this paper highlights some of these benefits in both a normal and generalised modelling framework, and examines the estimation of between-study correlation to aid practitioners. Traditionally, meta-analysis models combine summary measures of a single quantitative endpoint, taken from different studies, to produce a single pooled result. However, multiple pooled results are required whenever there are multiple outcomes  or multiOften the advantage of a multivariate random-effects meta-analysis lies in its ability to use the between-study correlation of the multiple endpoints of interest. For example, in diagnostic studies the sensitivity and specificity are usually negatively correlated across studies due to the use of different thresholds . Van HouSome recent articles have indicated the between-study correlation may often be estimated at the end of its parameter space, as +1 or -1. Thompson et al.  apply a The outline of the paper is as follows. We begin by introducing the general normal model for BRMA and discuss analytically why the between-study correlation can be estimated at the edge of its parameter space. We then apply the model to two real examples from the literature, one involving a diagnostic marker and one involving a surrogate outcome, and these both give a between-study correlation estimate of -1. This then motivates a realistic simulation study to examine how the estimate of between-study correlation affects the statistical properties of the pooled and between-study variance estimates. It also allows us to compare the performance of the normal BRMA model to two separate URMAs, the more common approach in practice . We thenIn this section we introduce a hierarchical normal framework for BRMA and URMA. We describe how the BRMA normal model is estimated, and analytically consider the estimation of between-study correlation. We then describe the rationale and methodology for our simulation study of BRMA versus URMA. To ensure a real world context, this section also includes two motivating examples from the medical literature where a BRMA is potentially important.Glas et al.  systemati = 1 to n studies are identified by a systematic review, and that two endpoints (j = 1 or 2) are available from each study. Each study supplies summary measures, Yij, and associated standard errors, sij, for each endpoint. For instance, for diagnostic studies Reitsma et al. . Cholesky decomposition of Ω also helps ensure convergence when B is very close to 1 or -1.In our analyses of equation (1) in this paper the between-study parameters  restrictions on V, namely that V - δ is a covariance matrix, that is non-negative definite. If the estimated V does not satisfy this restriction, the maximum likelihood estimate of Ω will be truncated on the boundary of its parameter space. This means that if Ω is diagonal (as for URMA) the maximum likelihood estimate on the boundary will have one or both of the between-study variances equal to zero; else if Ω is non-diagonal (as for BRMA) then either one or both of the between-study variances equals zero or else the between-study correlation equals -1 or +1. In meta-analysis a between-study variance estimate of zero is well-understood, but a between-study correlation estimate of -1 or +1 is likely to be less familiar to practitioners. Thompson et al. [V, and one that prevents a variance estimate < 0 or a correlation estimate > +1 or < -1, as might otherwise be obtained.Estimation and inference in classical linear mixed models are based on the marginal model, which for equation (1) is the bivariate normal model with variance-covariance n et al.  refer toB is likely to be +1 or -1 and examine the impact this has on the pooled and between-study variance estimates. We will also evaluate the benefits of BRMA over two separate URMAs, the more common approach in practice, and explore extensions to a generalised BRMA model for meta-analysis of proportions. To achieve these goals we firstly apply the normal BRMA of equation (1) to the two motivating examples. We then perform a simulation study of the normal BRMA model, as described below. The generalised BRMA model is then introduced and assessed in relation to the normal BRMA model and two separate generalised URMAs (see Results).In this paper, to aid practitioners we will further assess the normal BRMA model and the role of between-study correlation. In particular, we aim to identify the situations when n = 5 or n = 50 studies for complete data, and either n = 10 or n = 50 for missing data. Our method of simulation was deliberately chosen to be similar to that previously used by Berkey et al. [n = 50.We carried out a simulation study of the general normal model for BRMA in 11 scenarios, labelled (i) to (xi) Table . Each ofy et al. . As an eβ1 = 0 in order to reflect little clinical benefit (e.g. a sensitivity of 50%) and in contrast β2 = 2 (e.g. a specificity of 88%). The within study variances required, i.e. the 50 N distribution and squaring the value obtained. This produced a median We chose n et al.   were analysed separately by:ρB = 0) using REML to estimate β1, β2, • fitting two separate URMAs as in equation (2)  using REML to estimate The 1000 BRMA estimates and the corresponding 1000 URMA estimates from scenario (i) were then compared by calculating:• average parameter estimates across all the simulations (to check for bias)β1 and β2• coverage of the 95% confidence intervals for β1 and β2• average standard error and mean-square error (MSE) of B was equal to +1 or -1 in the BRMA• the number of occasions βj were calculated using:To assess coverage, the 95% confidence intervals for nj the number of studies providing endpoint j. This t-distribution is commonly used in the meta-analysis literature, although it is only an approximation [with ximation .n = 50 in scenario (viii) each of the 1000 simulated meta-analyses contained 25 studies with complete data and 25 studies with data for the first endpoint only.Simulations in the other scenarios followed in the same manner as described above but with the data generated from the parameter values specific to each scenario as given in Table ρB with the log-likelihood gradually increasing as B approaches -1 ) and then two separate URMAs (equation (2)) were applied to the telomerase data. Both approaches gave a pooled sensitivity of about 76% and a pooled specificity of about 88% Table . The BRM) and theρB at the boundary of its parameter space adversely influencing the other BRMA parameter estimates and, if so, how (e.g. does it introduce bias)? Also, in terms of the individual pooled estimates, the telomerase and CD4 examples indicate little benefit of BRMA over two separate URMAs, despite the utilisation of correlation; but is this generally true and in what situations should a BRMA be preferred? To understand the answers to these important questions, it is helpful to now consider the results from our simulation study.The question is thus posed: is the estimation of Ω, rather than using Cholesky decomposition of Ω as previously. Interestingly, this approach produced non-sensical between-study correlation estimates of -1.12 and -1.074 for the telomerase and CD4 datasets respectively. This emphasises the importance of a boundary constraint on ρB as imposed by the Cholesky decomposition.We note at this point that, for both telomerase and CD4, we also tried estimation of BRMA using an unstructured form of Table ρB, as either +1 or -1, especially when the number of studies in the meta-analysis is small. For example, with n = 5 in scenario (ii), where the true ρB was 0.8, 605 of the 1000 simulations (60.5%) gave B equal to +1 and 103 simulations (10.3%) gave B equal to -1. This led to a mean value of B equal to 0.639, a downward bias of about 20%. However, this downward bias is not in itself a concern because it is simply caused by the maximum likelihood estimator sensibly truncating B at +1 and -1, which improves the mean-square error of B. Furthermore, B is clearly asymptotically unbiased, with the occurrence of B equal to -1 or +1 and thus the bias in mean B becoming increasingly less as the number of studies in the meta-analysis increases , where the within-study variation was relatively large, 58% of the simulations gave a between-study correlation of -1 or +1.One can see from Table es Table . InteresB equals -1 in the BRMAs of the telomerase and CD4 datasets. For the telomerase data there are 10 studies; the simulations show that this magnitude of studies will often provide little information about ρB, causing B to often be constrained at -1 or +1 so that the restrictions imposed on V are met. For the CD4 data, even though there are five more studies than telomerase, the mean within-study variance for endpoint j = 1 is 0.15 and this is large relative to the between-study variation (B will often require truncation at the end of its parameter space to ensure V is non-negative definite.These findings indicate why n = 5 n = 5), the normal BRMA model produces a noticeable upward bias in the between-study variance estimates. For example, in scenario (ii) with n = 5 there was an upward bias of 0.024 in τ12) is formulated by τ12 = ρB τ1 τ2. Now, if B is constrained at -1 or +1, then to obtain the necessary solution for 12 the maximum likelihood estimator can only increase the j s, which do not have an upper bound constraint. Thus the between-study variance estimates are inflated to compensate for the constraint on B. This explains why the BRMAs of the telomerase and CD4 data, where B was truncated at -1, give j s that are noticeably larger than those from two separate URMAs. Practitioners need to be aware of this issue; however, we do not consider it a major concern as the maximum likelihood estimator for B. Furthermore, the inflation is essentially conservative, leading to a larger standard error and mean-square error of pooled estimates as now discussed.Our simulation results show that the between-study variance estimates were less frequently truncated at zero in the BRMA than the URMA Table . For exaB was often + 1 or -1 it is encouraging that, despite the upward bias in between-study variances, there was no systematic bias in the pooled estimates from the BRMA (Table n = 5 results for scenario (iii), where about 56% of simulations gave B equal to +1 or -1 and there was an upward bias in between-study variances, the standard error/mean-square error of 1 was larger in the BRMA (0.272/0.083) than the URMA (0.268/0.081). This explains the BRMA results for telomerase and CD4, where the standard errors of pooled estimates were larger in the BRMA than the URMA due to the inflated between-study variances.For all complete and missing data scenarios, the pooled estimates were approximately unbiased for both BRMA and URMA. Even in those scenarios where MA Table . The maiβ1 and β2 was between 93% and 98% in most scenarios assessed, and was often similar in URMA and BRMA. It is hard, though, to make general conclusions regarding coverage, as those situations where B is often +1 or -1 are the same situations where the t-distribution with n-1 degrees of freedom is a poor approximation to the true sampling distribution. The true degrees of freedom to use here are complex and account for the within-study variances [The coverage of ariances ; howeverB was often +1 or -1 as discussed above. We also compared the subset of BRMA results where B did not equal +1 or -1 with the corresponding URMA results, and again found that BRMA was generally slightly superior to URMA. This finding agrees with previous algebraic results [β1 and β2 themselves. Our focus here is on the individual pooled estimates, but we note that there are also broader reasons why a BRMA may be preferred over URMA for complete data. These are summarised in the Discussion to ensure a more complete picture for practitioners considering BRMA.For complete data, in most scenarios the BRMA was marginally superior to URMA as the pooled estimates had slightly smaller standard errors and mean-square errors, especially given large correlations Table . However results , that gij = 2 was of particular interest because of the missing data for this endpoint. Encouragingly, the mean-square error and mean standard error of 2 were much smaller in the BRMA than the URMA, although the coverage was comparable (Table n = 10 simulations of scenario (xi) the mean standard error was 0.225 in the BRMA compared to 0.262 in the URMA, and the MSE was 0.0708 in the BRMA compared to 0.0921 in the URMA. The reduction in standard error and MSE was larger when both the within- and between-study correlations were high. Even when ρWi was zero there was still a reasonable benefit if ρB was high; for example, in the n = 10 simulations of scenario (viii) the mean standard error of 2 was 0.174 in the BRMA compared to 0.209 in the URMA. This finding agrees with algebraic work regarding the benefits of BRMA for when there are data missing at random [For the missing data simulations, the pooled estimate for endpoint le Table . For exaIn our above simulations of the normal BRMA model we used non-negative within- and between-study correlations; however, in reality negative correlations may arise as in the telomerase and CD4 examples. Also, our simulations took the within-study correlations to be the same in each study, while in reality their value may vary. Further simulations were thus performed to assess negative correlation and discrepant within-study correlations. These gave findings consistent with those identified previously ; thus, our conclusions are only valid for Yij s that can truly be assumed normally distributed. This normality assumption is common in the meta-analysis field, and will often be suitable (see Discussion). However, in our two motivating examples it is more plausible for the CD4 data than the telomerase data as the latter involves a meta-analysis of proportions, for which the normality assumption is not appropriate when some studies have a small number of patients or the proportions are close to 0 or 1. For this reason recent articles [So far in this paper we have modelled the summary statistics across studies, i.e. the articles  suggest articles :i~Binomial  logit(sensitivityi) = β1 + u1no. testing positivei~Binomial  logit(specificityi) = β2 + u2no. testing negativeρB = 0), emphasising that the BRMA is also beneficial over URMA in the generalised model framework. Note though that, although it is the best method, the generalised BRMA model is itself not without bias (Table Equation (3) can be fitted using maximum likelihood estimation in SAS NLMIXED. Chu and Cole  show thaas Table ; to rect1 as 30.5, whereas for another set the standard error was close to zero. Indeed SAS provides the following warning: 'the final Hessian matrix is not positive definite, and therefore the estimated covariance matrix is not full rank and may be unreliable'. The problem here is again due to the between-study correlation of estimate -1 in Ω, as this causes the determinant of δi + n = 10 simulations of the generalised BRMA model . However, there are broader reasons why BRMA may still be preferable in this situation (see below).Our simulation results highlight that a normal model for BRMA is preferable to two separate URMAs for estimating the pooled endpoints, and our results are consistent with previous findings that show how the inclusion of correlation allows the 'borrowing of strength' across endpoints ,10,19. Wpersonal communication).In equation (3) we extended our work to a generalised BRMA model for meta-analysis of two proportions. For synthesis of two proportions, like sensitivity and specificity, this approach is preferable to the general normal BRMA model Table  because β1 - β2 [β1 + β2 [β1/β2 [β1 + β2)) [Our simulations focused mainly on the benefits of BRMA over URMA for estimating the pooled endpoints. However, there are also broader reasons why a BRMA may be preferable to URMA, for either complete or missing data. For example, BRMA allows one to describe the bivariate relationship between endpoints ,5, modelβ1 - β2 , β1 + β22 [β1/β2 . For ins1 + β2)) . Further1 + β2)) ,8, and f1 + β2)) . Berkey ρWi s to be taken into account, as in practice they will only be estimates themselves as mentioned by Daniels and Hughes [personal communication).For the general normal BRMA model, the role of estimation techniques other than REML would be interesting to consider, especially as other potentially better options have just been proposed . For thed Hughes . HoweverThe use of individual patient data (IPD) in multivariate meta-analysis should also be considered, especially as IPD is the gold-standard for meta-analysis  and it wIn this paper we have used analytic reasoning, two applied examples and a realistic simulation study to highlight the benefits of a normal model for BRMA over two separate URMAs, and explain why the between-study correlation is often estimated as +1 or -1. For meta-analysis of proportions, we also extended our work to a generalised model for BRMA, to ensure the binary data is modelled correctly. Our work adds to a growing body of literature indicating the rationale and benefits of a multivariate approach to meta-analysis, and we encourage meta-analysts to consider the approach in practice.BRMA – Bivariate random-effects meta-analysisIPD – Individual patient dataURMA – Univariate random-effects meta-analysisREML – Restricted maximum likelihoodThe author(s) declare that they have no competing interests.RDR, KRA, AJS and PCL developed the study. RDR performed analysis of the telomerase and CD4 data. RDR, PL and JRT performed the normal model simulations; RDR performed the generalized model simulations. All authors examined the simulation results and discussed their interpretations. RDR wrote the paper with contributions from all authors.The pre-publication history for this paper can be accessed here:Appendix 1. Simulation results of the normal BRMA and URMA models for scenarios (iii) to (vii), (x) and (xi)Click here for fileAppendix 2. Simulation results for the normal BRMA and URMA models for some scenarios involving negative correlationClick here for file


The electrical activity of the uterine muscle is representative of uterine contractility. Its characterization may be used to detect a potential risk of preterm delivery in women, even at an early gestational stage.We have investigated the effect of the recording electrode position on the spectral content of the signal by using a mathematical model of the women's abdomen. We have then compared the simulated results to actual recordings. On signals with noise reduced with a dedicated algorithm, we have characterized the main frequency components of the signal spectrum in order to compute parameters indicative of different situations: preterm contractions resulting nonetheless in term delivery  and preterm contractions leading to preterm delivery (i.e. high-risk contractions). A diagnosis system permitted us to discriminate between these different categories of contractions. As the position of the placenta seems to affect the frequency content of electrical activity, we have also investigated in monkeys, with internal electrodes attached on the uterus, the effect of the placenta on the spectral content of the electrical signals.In women, the best electrode position was the median vertical axis of the abdomen. The discrimination between high risk and normal contractions showed that it was possible to detect a risk of preterm labour as early as at the 27th week of pregnancy (Misclassification Rate range: 11–19.5%). Placental influence on electrical signals was evidenced in animal recordings, with higher energy content in high frequency bands, for signals recorded away from the placenta when compared to signals recorded above the placental insertion. However, we noticed, from pregnancy to labour, a similar evolution of the frequency content of the signal towards high frequencies, whatever the relative position of electrodes and placenta.On human recordings, this study has proved that it is possible to detect, by non-invasive abdominal recordings, a risk of preterm birth as early as the 27th week of pregnancy. On animal signals, we have evidenced that the placenta exerts a local influence on the characteristics of the electrical activity of the uterus. However, these differences have a small influence on premature delivery risk diagnosis when using proper diagnosis tools. H: fast wave high) and a low (FWL: fast wave low) frequency components  for recording and a reference electrode was positioned on the hip of the patients. We used a device specifically developed for EHG recordings [This study obtained the agreement of our regional ethical committee. Following informed consent, 111 recordings were made from 107 women hospitalized for risk pregnancy in the Amiens Hospital. Inclusion criteria were singleton pregnancy, term between 18 and 37 weeks of pregnancy and the presence of uterine contractions. Only one third of the women finally delivered preterm. Term labour was defined as delivery occurring after the 37d between the median vertical axis of the abdomen and the electrodes, as well as the influence of the inactive tissue depth (ITD) of pregnant women. We have modified for this purpose the equation proposed by Farina et al. for the conductive volume used to model the striated muscle surface EMG [d = 5 cm and we compared these distant signals with the reference one (d = 0) by using Kruskall-Wallis and Mann-Whitney statistical tests.The first requirement to monitor uterine activity is to obtain good quality signals. We have studied the influence of the position of the recording electrodes with two approaches. The first one was to use a mathematical model of the filtering effect of the tissues present between the uterus and skin electrodes Figure . These tface EMG . We alsoL and FWH components by using wavelet transform [Since several noises  corrupt the recorded signals, and therefore modify the frequency content of raw EHG signals, we used a dedicated algorithm, previously developed in the laboratory , to reduransform . We compransform .The last step was decision making. We compared different neural networks algorithms, which are non-linear classification tools, to test the power of EHG analysis to diagnose a risk of preterm birth. These neural networks, by using the parameters extracted from the EHG bursts, permitted us to classify, for a given stage of gestation (i.e. 27–28 weeks of gestation), the contractions between two groups: i) high risk contractions: contractions recorded on women who proved to deliver preterm (i.e. at 33 weeks of gestation), ii) normal contractions: contractions recorded on women who proved to deliver at term (i.e. at term > 37 weeks of gestation). We used Multi Layer Perceptrons with one hidden layer and supervised learning. Cross-validation with a classified subset was used for the learning phase. We tested two different learning algorithms. The first one is based on the gradient back-propagation with momentum (BPM)  and the Internal EMG signals were obtained from three pregnant cynomolgus monkeys by the implantation of a telemetric transducer TL 11M3-D70-CCP  having specific filters (1st order high-pass Bessel filter (0.1 Hz), 5th order low-pass Bessel filter (100 Hz)). The transducer contains two bipolar EMG leads, which were sutured on the uterine wall, and one pressure catheter, which was used to measure the intra uterine pressure (IUP). The transducer was implanted chronically at 120 days of gestation and remained until labour. We sutured one EMG bipolar lead above the placenta (P) and the other away from the placental implantation site (NP). Our Institutional Animal Use and Care of Laboratory Research Animal Committee approved the procedures. Signals were digitized with a 50 Hz sampling rate. Contractions were extracted off line by a peer review of the signals and distributed according to three gestation stages , by analysing the number of contractions per 10 minutes, and the amplitude of the intra-uterine pressure. For each contraction, we calculated the relative energy spectrum (RES) of the associated EMG.d and ITD influences separately and then both influences together. ITD influences the very high frequency content of the EHG spectrum and d influences the median and high part of the spectrum  Table  and of sH increases, subsequently associated with a decrease in FWL relative energy.Figure H towards lower frequencies when the electrodes are located above the placental (P) insertion. However, we evidenced the same global shift of the content of the EMG signal towards higher frequencies, as pregnancy progresses, whatever the electrode position was with respect to the placenta insertion.Figure Furthermore, if we study the placental influence on EMG associated with labour contractions, the differences between "No Placenta" and "Placenta" EMG RES is decreased (data not shown).H frequency band.The first step to monitor biomedical signals is to record them in the best conditions. For the EHGs recorded on women, our study proved that the best way to position the electrodes, as far as the frequency content of the EHG is concerned, is to align the electrode pairs with the median vertical axis of the uterus. In this location, the electrodes are closer to the uterus and the spectral content of the EHG presents higher energy in the FWConcerning the diagnosis of risk contractions, a problem remains for the still limited pertinence of our results because at best, the results of the LM algorithm ranges are: MR = 11%–19.5%, Se = 0.79–0.82, Sp = 0.87–0.88, NPV = 0.85–0.87. One explanation is the low number of contractions available for learning, compared to the number of parameters that were fed into the learning system. In line with our previous results , we chosH) seems therefore to be of great importance to characterise contraction efficiency, thereby justifying the need for efficient recording and filtering techniques [The animal study has allowed us to show evidence for the changes along pregnancy, of the spectral uterine EMG content. The shift of the uterine EMG spectrum towards higher frequencies, when the date of the term approaches, has already been mentioned in previous experiments -4. We rechniques  of uteriWe have developed suitable signal processing tools to characterize the uterine contractions detected by their electrical activity during pregnancy, in monkeys and in the human. On animal signals, we have shown that the placenta exerts a local influence on the characteristics of the EMG, which changes along pregnancy. Proper diagnosis tools permit us to elucidate the diagnosis bias of the placental influence on the signals. On human recordings, we have proved that it is possible to detect the risk of preterm birth (identification of high risk contractions) by using EHG processing, at stages as early as at 27 weeks of pregnancy, and therefore to make an early diagnosis of preterm labour from the analysis of abdominal EHG, irrespective of where the placental insertion is. Collectively, these works allowed us to determine a reliable procedure to record properly the electrical uterine activity from the abdomen of pregnant women.BPM, back propagation with momentum algorithm; EHG, electrohysterograms; EMG, electromyograms; ITD, inactive tissue depth; LM, Levenberg-Marquardt algorithm; MR, Misclassification Rate; NP, extra-placental site of the uterus; NPV, Negative Predictive Value; P, placental site of the uterus; RES, relative energy spectrum; Se, Sensitivity; Sp, Specificity.The authors declare that they have no competing interests.CM has been involved in the conception and design of the recording protocols, in the analysis and interpretation of data as well as in drafting the manuscript. JT has been involved in the conception and design of the recording protocols, in human and animal signal recordings, in the design of software for analysis and interpretation of data. He also revised the manuscript. SR has been involved in the development of the mathematical model as well as for analysis and interpretation of data. She also revised the manuscript. GG has been involved in the conception and design of the animal recording protocol, in the analysis and interpretation of data as well as in revising the manuscript. All the authors have given final approval of the version to be published.


The success of a Fontan circulation depends on several factors including low pulmonary vascular resistance. Pulmonary vascular resistance rises in response to hypoxia. Hypoxia is associated with altitude. Therefore, we wondered whether altitude is a risk factor for early failure after the Fontan operation. The aim was to test this hypothesis.Data were obtained from all published series of 'total cavopulmonary' Fontan operations since 1990. The early failure rate from each series and the altitude of the respective cities were recorded. Early failure was defined as death, takedown of Fontan, or transplantation during the same hospital admission. The association between altitude and failure rate was investigated by rank correlation and logistic regression.24 series were identified from centres situated at altitudes ranging from sea level to 520 metres. The plot of failure rate versus altitude suggests that failure rate increases with altitude. Logistic regression did not fit the data adequately. This was possibly due to the influence of unmeasured and unknown factors affecting the results, as well as the fact that centres were not randomly chosen but were self-selected by virtue of publishing their results. However, Spearman's rank correlation was 0.74 (p = 0.001).The early outcome of the Fontan circulation appears to be adversely affected by altitude. Low pulmonary vascular resistance (PVR) is a major criterion for the successful creation of a Fontan circulation . It is aTherefore, we wondered whether altitude is a risk factor for early failure after completion of a Fontan circulation. The aim of this study was to test this hypothesis.The data of all published series on the Fontan operation obtained from a literature search were reviewed. For each series, the number of patients treated, the primary diagnoses (case mix), the exact technique of the Fontan operation, the city where the procedures were performed and the early failure rate were recorded. Early failure was defined as death, takedown of the Fontan pathway, or heart transplantation during the same hospital admission. The altitude of the publishing centres, measured in metres above sea level and recorded to the nearest 5 metres, were obtained from the Royal Geographical Society and from the website .Series that included patients operated on before 1990 were excluded for two reasons. The first was to reduce the impact of era on results. The second was to standardise the operations, since the procedure was still undergoing technical modifications in the 1980s. Therefore, only series incorporating the 'modern' Fontan completion techniques of total cavopulmonary connection by either lateral tunnel  or extraThe relationship between early failure rate of the Fontan circulation and altitude was analysed using the Spearman rank correlation test and binary logistic regression as implemented in the Egret statistical package .24 series were identified to fulfil the inclusion and exclusion criteria -30. TheyOne series exclusively dealt with hypoplastic left heart syndrome, and the right ventricle supported the systemic circulation . In the Figure The concern that altitude may have a negative impact on the outcome of the Fontan procedure is not new. Workers at Salt Lake City investigated the impact of altitude of the patients' home on long-term outcome . They foBoth atmospheric pressure and partial pressure of oxygen fall even with a modest rise in altitude  and thisSome limitations of this study must be recognised. We have tried to minimize bias due to surgical technique by adhering to the contemporary series with comparable techniques of lateral tunnel or extra cardiac conduit Fontan circulations. These two techniques have similar outcomes according to the available evidence. However, several limitations persist. Although series do report the rate of Fontan takedown, it is not always clear whether this includes those that were taken down at the time of the original operation. Such a situation is rare, but it is possible that a surgeon may make such a decision when faced with problems and challenges at the time of surgery suggestive of early post-operative failure. It is possible that such cases may not be regarded as completed Fontan procedures and are therefore not included in the series. Clearly, this would result in under-reporting of the true failure rate. There is slight variation between the series as to whether the completion of the circulation was performed as a single or two-stage procedure but there is an even greater variation in the practice of fenestration of the Fontan conduit. These factors may bias the results. This study takes no account of patient selection, the skill and experience of the surgeon, the experience of the centre and the methods of post-operative management as these issues are rarely addressed in the published series. There are only 24 centres in this study and even fewer at higher altitudes. These may not be representative of all high altitude centres. Indeed, centres in this study were not randomly chosen but were self-selected by virtue of having published their Fontan results. Finally, pre-operative PVR of these patients prior to completion of the circulation is not available and therefore analysis of the relationship between PVR and altitude could not be performed.Although the above limitations dictate a certain level of caution in the interpretation of the results, the data in this study suggest that the success of establishing a Fontan circulation may indeed be sensitive to altitude.


The evidence on prematurity as 'a priori' a risk for palatal disturbances that increase the need for orthodontic or orthognathic treatment is still weak. Further well-designed clinical studies are needed. The objective of this review is to provide a fundamental analysis of methodologies, confounding factors, and outcomes of studies on palatal development. One focus of this review is the analysis of studies on the palate of the term newborn, since knowing what is 'normal' is a precondition of being able to assess abnormalities.A search profile based on Cochrane search strategies applied to 10 medical databases was used to identify existing studies. Articles, mainly those published before 1960, were identified from hand searches in textbooks, encyclopedias, reference lists and bibliographies. Sources in English, German, and French of more than a century were included. Data for term infants were recalculated if particular information about weight, length, or maturity was given. The extracted values, especially those from non-English paper sources, were provided unfiltered for comparison.The search strategy yielded 182 articles, of which 155 articles remained for final analysis. Morphology of the term newborn's palate was of great interest in the first half of the last century. Two general methodologies were used to assess palatal morphology: visual and metrical descriptions. Most of the studies on term infants suffer from lack of reliability tests. The groove system was recognized as the distinctive feature of the infant palate. The shape of the palate of the term infant may vary considerably, both visually and metrically. Gender, race, mode of delivery, and nasal deformities were identified as causes contributing to altered palatal morphology. Until today, anatomical features of the newborn's palate are subject to a non-uniform nomenclature.Today's knowledge of a newborn's 'normal' palatal morphology is based on non-standardized and limited methodologies for measuring a three-dimensional shape. This shortcoming increases bias and is the reason for contradictory research results, especially if pathologic conditions like syndromes or prematurity are involved. Adequate measurement techniques are needed and the 'normal palatal morphology' should be defined prior to new clinical studies on palatal development. The proThe factors discussed as potential triggers of a premature birth include: high or low age of the mother, low socio-economic status, inadequate antenatal care, drug, alcohol and nicotine abuse, diabetes, multiple pregnancies , anemia,Preterm infants suffer not only from the effects of a shorter antenatal development period but from the immaturity of their organs , involvith – 26th gestational week. Although substantial health impairment is to be expected in 2/3 to 3/4 of infants born in the 24th gestational week, the survival rate is meanwhile 50 – 60%, and in those born in the 25th GW as high as 70 – 80%  preterm infant, [BW] birthweight, [LBW] low birthweight, [NBW] normal birthweight, [VLBW] very low birthweight, [NBW] normal birthweight, [GA] gestational age, [GW] gestational weeks, [NS] not significantThe author(s) declare that they have no competing interests.AH designed the study, searched the databases, extracted the data, analyzed the results and wrote the manuscript. HR helped with study design, analysis and provided critical input in neonatal associated issues and revised the manuscript. UE and EH formulated the research question, helped with study design, analysis and in revising the manuscript. All authors read and approved the final manuscript.Table 1. Excluded studies and reasons for exclusion.Click here for fileTable 3. Studies on plaster casts.Click here for fileTable 2. Nomenclature of palatal structures as presented in Figure Click here for file


GRL) variants with obesity. Previous studies have linked the N363S variant of the GRL gene to increased glucocorticoid effects such as higher body fat, a lower lean-body mass and a larger insulin response to dexamethasone. However, contradictory findings have been also reported about the association between this variant and obesity phenotypes. Individual studies may lack statistical power which may result in disparate results. This limitation can be overcome using meta-analytic techniques.Since both excess glucocorticoid secretion and central obesity are clinical features of some obese patients, it is worthwhile to study a possible association of glucocorticoid receptor gene . Genotype was assessed by PCR-RFLP (Tsp509I). The final formal meta-analysis included a total number of 5,909 individuals.We conducted a meta-analysis to assess the association between the N363S polymorphism of the 2 (95% CI: +0.004 to +0.35) for homo-/heterozygous carriers of the 363S allele of the GRL gene in comparison to non-carriers. Moreover, differences in pooled BMI were statistically significant and positive when considering one-group studies from the literature in which participants had a BMI below 27 kg/m2 (+ 0.41 kg/m2 [95% CI +0.17 to +0.66]), but the differences in BMI were negative when only our novel data from younger (aged under 45) and normal weight subjects were pooled together (-0.50 kg/m2 [95% CI -0.84 to -0.17]). The overall risk for obesity for homo-/heterozygous carriers of the 363S allele was not statistically significant in the meta-analysis .The meta-analysis revealed a higher body mass index (BMI) with an overall estimation of +0.18 kg/mGRL gene is associated with either average BMI or obesity risk.Although certain genotypic effects could be population-specific, we conclude that there is no compelling evidence that the N363S polymorphism of the  According to the updated 2004 version of the Human Obesity Map, 113 candidate genes showed initial association with obesity, out of which only 18 genes are supported by at least five positive studies ) were considered separately. However, the pooled difference in BMI was negative and statistically significant in homo-/heterozygous carriers of the S363 allele when only novel data for normal weight subjects were examined . Opposite tendencies for differences in BMI were also found when performing the meta-analysis after stratifying by gender, prevalence of the polymorphism and considering children and adult populations, separately  and their 95% confidence intervals (CIs) to further explore the association between the N363S polymorphism of the  Figures , 2, 3. TGRL gene on the incidence of obesity. Although procedures of meta-analyses have mostly been applied to comprehensively examine the results from previously reported studies, we here also include novel data from three studies derived from Spanish and German populations of our group that have not been published so far. Moreover, the meta-analysis was performed only after relevant data that were not available in the original publications were obtained contacting directly with the authors. The new data set was generated using two different strategies: the first was to match controls to cases by gender and age, whereas the second one was performed as a case-parent study that can be considered as a special type of case-control study in which cases (unrelated probands) and pseudocontrols  are matched by ethnic origin. The magnitude of the association can be assessed by odds ratios (OR) in the case-control study, while the deviations of the transmission probabilities from the expected value under the null hypothesis of no-association (50%) are related to the genotype relative risk , after adding our own new data to previously published results, data from 5,909 unrelated subjects from a diverse ethnic backgrounds were pooled together for the analysis of BMI differences between homo-/heterozygous carriers and non-carriers of the 363S allele of the GRL gene. The overall pooled estimated genotype frequency for homo-/heterozygous carriers of the 363S allele was 4.47% (95% CI: 4.24–5.30) with a substantial heterogeneity (p < 0.001). Interestingly, for BMI differences, the overall estimation was +0.18 kg/m2 (95% CI: +0.004 to +0.35) higher in homo-/heterozygous carriers of 363S allele of the GRL gene than in non-carriers, also with an important heterogeneity (p < 0.001). It is worthwhile to mention that the pooled analysis (i.e. the "one population studies") revealed a large range for BMI: from non-obese  to severely-obese individuals  and also with regard to age: from 18 years . It also yields a much better opportunity for the detection of significant effects in sub-group analyses of homogeneous data. Thus, the meta-analysis represents a useful strategy to assess the potential contribution of variants with rather low allele frequencies (i.e. the N363S polymorphism of the GRL from 4 to 9%), which may exert only moderate effects on polygenic disorders.Overall, our analyses did not indicate an increased risk for the development of obesity for homo-/heterozygous carriers of the S363 allele of the GRL gene is associated with the average BMI or with an increased prevalence of obesity, in the large number of human studies included in our analyses.Although certain genotypic effects could be population-specific, we conclude that there was no compelling evidence that the N363S polymorphism of the The author(s) declare that they have no competing interests.AM participated in the design of this study, recruited the meta-analysis data, contributed to the analysis and interpretation of the data, was responsible for Spanish subject enrolment and draft the manuscript. MCO carried out the laboratory analyses, participated in the analysis and interpretation of the data and helped to draft the manuscript. ASV made the statistical analysis and figures. JAM ad MAMG participated in the design of this study and in the analysis and interpretation of the data. JH, AH and HV participated in the initiation and design of this study and in the analysis and interpretation of the data and were responsible for German subjects enrollment. All authors read and approved of the final manuscript.The pre-publication history for this paper can be accessed here:


DRD2 TaqI A and DRD4 48 bp VNTR genetic polymorphisms and performance on a behavioral measure of impulsivity, the delay discounting task (DDT), and three traditional self-report measures.Research on the genetic basis for impulsivity has revealed an array of ambiguous findings. This may be a result of limitations to self-report assessments of impulsivity. Behavioral measures that assess more narrowly defined aspects of impulsivity may clarify genetic influences. This study examined the relationship between possession of the DRD2 TaqI A locus, individuals were designated as possessing at least one copy of the A1 allele (A1+) or not (A1-), and for the DRD4 48-bp VNTR locus, individuals were designated as having at least one long allele  or not (L-). Principal analyses used multiple univariate factorial 2 (A1+/A1-) × 2 (L+/L-) analyses of variance.195 individuals  were recruited from a university campus and were assessed in small group sessions using personal computers. Genotyping was conducted using previously established protocols. For the p = .006) as well as a significant interaction effect (p = .006) between both genes. No other significant effects were evident on the self-report measures, with the exception of a trend toward an interaction effect on the Sensation Seeking Scale. Exploratory analyses suggested that the significant effects were not a function of population stratification or gender.A significant main effect of A1+ status on DDT performance was evident (DRD2 TaqI A and DRD4 VNTR polymorphisms influence impulsivity as measured with a delay discounting task. Specifically, these findings suggest that an interaction between the functional effects of the two unlinked genotypes results in significant difference in the balance of mesolimbic dopaminergic activation relative to frontal-parietal activation. However, these findings are also the first in this area and must be replicated.These data suggest that the DRD2 TaqI A and DRD4 VNTR polymorphisms in the expression of impulsivity and provide initial support for the utility of using behavioral measures for clarifying genetic influences on impulsivity.These findings suggest a meaningful interaction between the  The effort to characterize the behavioral effects of genetic polymorphisms has produced a massive web of ambiguous associations and linkages -3. One sImpulsivity is a prototypic candidate for the endophenotype approach because it is a trait that varies considerably in the overall population -12 and i5HT2A receptor – G-1438A polymorphism and increased impulsivity, but Patkar et al.  and  and 58]]The most widely studied behavioral measure of impulsivity is the delay discounting task (DDT). From a delay discounting perspective, impulsivity is defined as the relative preference for a smaller reward, sooner in time, compared to a larger reward, later in time ; that isVersions of the delay discounting paradigm may also be used to study impulsivity in animal models ,59,60,73Given the limitations to the current literature on impulsivity as an endophenotype and the potential promise of using behavioral measures, in the current study we examined impulsivity as a potential endophenotype using two dopaminergic genetic polymorphisms as candidates for observed variation in impulsivity as measured by the DDT and three traditional measures of impulsivity. These three measures include the BIS, Eysenck Impulsivity Questionnaire (EIQ), and the Sensation Seeking Scale – Form A (SSS), all of which have undergone extensive psychometric validation ,12,76. TDRD2 TaqI A and DRD4 48 bp VNTR polymorphisms. Both have been associated with psychiatric disorders involving impulsivity, namely substance abuse and ADHD . In c system -87, the c system -61,73-75DRD2 TaqI A site is a SNP with two possible alleles, the major A2, and minor A1. The A1+ genotype (heterozygous or homozygous A1) has been most strongly associated with substance abuse, particularly alcoholism , we alsA total of 195 unselected subjects were recruited from the Human Subject Research Pool at the State University of New York at Binghamton and are described demographically in Table DRD2 TaqI A was typed with a PCR/RFLP method , fo, fo120]]ypes see . Correctypes see , and becypes see .R2 = 0.904, interquartile range [IQR] = .72 – .96), which is comparable to past studies [R2 ≤ .30), and were excluded from consideration in the primary DDT analyses. This further increased the median R2 value for the remaining 166 subjects . As anticipated, subjects' performance on the DDT topographically resulted in hyperbolic discounting curves, exhibiting precipitous initial discounting followed by more modest decreases based on delay  than L+ (0%) for only the BIS Motor Impulsivity scale (Fishers Exact p = .032). Pearson's product-moment correlations between the indices of impulsivity are shown in Table Between 2.6% and 13.8% of values were missing from each of the remaining self-report scales because subjects skipped one or more items or incorrectly entered their subject ID number Table . There wDRD2 A1+/A1-) × 2 (DRD4 L+/L-) ANOVAs were run for each impulsivity index. There was a main effect of DRD2 TaqI A genotype  = 7.65, p = .006, ηp2 = .046), reflecting greater delay discounting measured impulsivity in A1+ individuals, and an interaction effect between the two genotypes  = 7.63, p = .006, ηp2 = .046), such that A1+/7R+ individuals exhibited the greatest delay discounting. Both effects are indicated in Table ps = .001–.007), but the other three groups were not significantly different from each other (ps = .241 – .968).Factorial 2  = 3.52, p = .06, ηp2 = .02). This effect also reflected greater sensation seeking in the A1+/L+ group, and is depicted in Table No significant effects were evident for any of the self-report measures of impulsivity, although a marginally significant interaction effect of k values  = 1.556, p = .215, ηp2 = .027). Median k values (IQR) across groups accounting for at least 10% of the sample were: European, .0416 (.0140–.1600); East Asian, .0349 (.0112–.1505), and Latin American, .0727 (.0133–.2198). Repeating the principal analyses in the largest single racial group, those with European ancestry, the main effect of DRD2 on DDT k values was marginally significant  = 3.487, p = .066, ηp2 = .047), as was the interaction effect of DRD2 and DRD4 on DDT k  = 2.729, p = .103, ηp2 = .038). However, the ηp2 effect sizes were very similar between the analysis of all subjects and European ancestry subsample, suggesting that the reduction of statistical significance was related to reduced statistical power in the smaller subsample. Taken together, these analyses suggest that the significant effects were not the result of population stratification.We explored the possibility of population stratification using the proposed approach. A one-way ANOVA revealed no significant effect of racial group on DDT F  = 13.448, p < .001, ηp2 = .071; BIS Total F  = 6.570, p = .011, ηp2 = .040; EIQ Venturesomness: F  = 8.924, p = .003, ηp2 = .047; EIQ Impulsivity: F  = 4.499, p = .035, ηp2 = .025; SSS Boredom Susceptibility: F  = 4.731, p = .038, ηp2 = .024) with males scoring higher in each case, which was not an unexpected sexual dimorphism [DRD2 TaqI A effect and interaction effect on DDT performance remained significant with the inclusion of gender as an independent variable  = 8.585, p = .004, ηp2 = .052; Interaction: F  = 7.683, p = .006, ηp2 = .047), further bolstering the stability of the finding. All other main and interaction effects were non-significant (not shown). Finally, to explore whether the use of a model fit criterion affected the DDT findings, the data were re-analyzed using all k values. This resulted in the same DRD2 TaqI A main effect  = 9.668, p = .002 ηp2 = .051) and interaction effect  = 4.830, p = .029, ηp2 = .026).Potential gender effects were examined by including gender as an additional two-level independent variable, finding several significant main gender effects , choices for immediate smaller rewards is associated with greater peak activation in the ventral striatum and regions of the medial prefrontal cortex and less relative activation in regions of the lateral prefrontal and parietal cortex [2 polymorphism may be modifying more primitive limbic neural systems involved in reward salience while the D4 polymorphism may be modifying higher-level systems involved in abstract thinking, deliberation and behavioral inhibition. Thus, particularly steep delay discounting (increased impulsivity) in those with both long DRD4 and A1 DRD2 alleles may reflect the concurrent variation in two key points in the reward decision-making loops spanning the corticostriatal-mesolimbic axis. Specifically, the pattern of findings suggest that for those individuals who have a greater sensitivity to reward based on possession of the DRD2 A1 allele, decreased frontal-cortical inhibition resulting from possession of a DRD4 VNTR long allele results in substantially greater discounting of delayed rewards.With regard to the mechanisms of underlying the interaction between the two loci, any discussion must be speculative given that the two polymorphisms under consideration have not been definitively characterized. Acknowledging this ambiguity, the observed associations may be understood in the context of the relative neuroanatomical localization of the Dl cortex . In contl cortex . Other sl cortex . TogetheDRD2 TaqI A main effect, the absence of a DRD4 VNTR main effect, and a significant interaction suggests that the influence of A1+-mediated functional variation in striatal D2 receptors has primacy over the influence of L+-mediated cortical D4 receptor variation in modulating inter-temporal choice. That is, for individuals with a mesolimbic dopamine system that is predisposed to favor dopaiminergic rewards (A1+), the presence of an allele that may be associated with reduced frontal cortical inhibitory control (L+) may result in disproportionately greater discounting of delayed rewards, but not the other way around. In cognitive terms, the results suggest that the effects of inhibitory considerations of the future are superimposed on the limbic signals of the incentive salience of immediate rewards. However, we reiterate that given that the functional roles of both genetic polymorphisms are far from fully understood, this explanation must be speculative at this point. Future studies using neuroimaging techniques may directly address this interpretation.Of interest, however, the contrapositive relationship by genotype was not indicated. Considered together, the presence of a Assuming these findings are genuine and can be replicated, they have potential implications for the ambiguity in association studies of genetic variables in psychiatric disorders -3. To daThere are a number of qualifications of these findings worth noting. Although they provide preliminary support for the notion that behavioral aspects of impulsivity may be more amenable to investigations of genetic influences, to our knowledge they represent the first report of specific genetic influences on delay discounting. As such, they must be replicated to affirm their empirical validity and to more conclusively affirm the potential of a task-based approach to generate more reliable findings than self-report measures. In addition, it should be noted that these findings most clearly apply to the European ancestry subsample, which both represented the largest proportion of subjects and was sufficiently large to affirm the findings independent of other ethnic subsamples. As such, although we found no evidence of population stratification, studies of delay discounting as a potential endophenotype in larger samples of non-European descendants will be important to address the generalizability of these findings.DRD2 TaqI A and DRD4 VNTR genes. Although these findings must be replicated, they may contribute to understanding the genetic basis of impulsivity and provide support for the notion of using behavioral tasks in doing so.In summary, the current study sought to examine the influences of two genetic polymorphisms on impulsivity using a behavioral task and traditional self-report measures. The findings indicated both independent and interactive genetic effects on the behavioral task, suggesting that genetically-based functional differences in the corticostriatal-mesolimbic dopamine system are responsible for variation in impulsivity from a delay discounting perspective. Rather than being a function of a single locus, these data suggest that the most prominent effect was as result of an interaction between polymorphisms of the DTAE participated in genotyping and phenotyping of subjects, study design, coordination, statistical analysis, and drafting the manuscript.JM conceived the study, participated in study design, coordination, statistical analysis, and drafting the manuscript.MM participated in genotyping and phenotyping of subjects and study design and coordination.JB participated in genotypingDD participated in genotypingSAL participated in study design.JKL oversaw genotyping, and participated in study design and coordination.DSW oversaw the study design and coordination.DRD2 TaqI A and DRD4 48 bp VNTR genotyping protocolsClick here for fileDelay discounting protocolClick here for fileAlternative DRD4 parsing method influences on DRD2 × DRD4 ANOVA to predict impulsivity phenotypes. This supplement explores parsing DRD4 using three alternative methods, and finds similar results to that of the main manuscript.Click here for file


Stavudine continues to be widely used in resource poor settings despite its toxicity. Our objective was to determine association between plasma stavudine concentrations and lipoatrophy, concentrations of glucose, lactate and triglycerides.Participants were enrolled in a cross-sectional study with lipoatrophy assessment, oral glucose tolerance test, fasting triglycerides, finger prick lactate, and stavudine concentrations. Individual predictions of the area under the concentration curve (AUC) were obtained using a population pharmacokinetic approach. Logistic regression models were fitted to assess the association between stavudine geometric mean ratio > 1 and impaired fasting glucose, impaired glucose tolerance, hyperlactataemia, hypertriglyceridaemia, and lipoatrophy.2 and 0.85 respectively. The median duration on stavudine treatment was 14.5 months. The prevalence of lipoatrophy, impaired fasting glucose, impaired glucose tolerance, hyperlactataemia, and hypertriglyceridaemia were 34%, 19%, 4%, 32%, and 23% respectively. Estimated median (interquartile range) stavudine AUC was 2191 (1957 to 2712) ng*h/mL. Twenty two participants had stavudine geometric mean ratio >1. Univariate logistic regression analysis showed no association between stavudine geometric mean ratio >1 and impaired fasting glucose (odds ratio (OR) 2.00, 95% CI 0.44 to 9.19), impaired glucose tolerance , hyperlactataemia , hypertriglyceridaemia , and lipoatrophy .There were 47 study participants with a median age of 34 years and 83% were women. The median body mass index and waist:hip ratio was 24.5 kg/mThere was a high prevalence of metabolic complications of stavudine, but these were not associated with plasma stavudine concentrations. Until there is universal access to safer antiretroviral drugs, there is a need for further studies examining the pathogenesis of stavudine-associated toxicities. Stavudine is no longer recommended as part of first line combination antiretroviral therapy (ART) because of a high cumulative risk of toxicity, notably symptomatic hyperlactataemia/lactic acidosis, lipoatrophy, and peripheral neuropathy ,2. In adIn 2006, WHO recommended reduced doses of stavudine following the findings of a systematic review that lower doses caused less toxicity without reducing efficacy ,5. Most,We investigated whether there was an association between stavudine plasma concentrations and lipoatrophy or concentrations of glucose, lactate and triglyceride in a population where stavudine use is likely to be widespread in the medium term: African HIV-infected adults.We conducted a prospective cross sectional study between February 2007 and January 2008. Ambulatory HIV-infected African black adults who presented for a routine follow up visit at public sector antiretroviral clinics in Cape Town were recruited by convenient sampling. Participants were eligible if they were on stavudine-based therapy for a minimum of 6 months. Participants with renal or hepatic disease, active opportunistic infections, known diabetes or dyslipidaemia, or self-reported non-adherence were excluded. All participants gave informed consent. The University of Cape Town research ethics committee approved the study.® lactate meter . Hyperlactataemia was defined as a lactate concentration greater or equal to 2.5 mmol/L.Participants fasted overnight and underwent an oral glucose tolerance test (OGTT). Impaired fasting glucose (IFG), impaired glucose tolerance (IGT) and diabetes were defined according to the American Diabetes Association criteria . FastingLipoatrophy was determined by self-reported peripheral fat loss using a validated questionnaire . LipoatrSelf reported adherence was determined using a standard 4-day adherence questionnaire administered by trained field workers . We reviWe measured plasma stavudine concentrations at 0, 30, and 120 minutes of the OGTT. We collected the blood samples using heparinised tubes that were immediately placed on ice until centrifugation within 4 hours, and then kept in a minus 80°C freezer until analysis. Stavudine was assayed by liquid chromatography tandem mass spectrometry using a validated method on an API 4000 mass spectrometer. The mobile phase consisted of gradient of acetonitrile and 0.5% glacial acetic acid. Chromatography was performed on a Phenomonex Synergi fusion C18 column maintained at 25°C. Reserpine was used as an internal standard. 50 μL of each sample was precipitated with acetonitrile containing the internal standard, centrifuged and 5 μL of the supernatant injected onto the column. Standard curves in the range 0.02 - 6 μg/mL and appropriate quality control samples were run with each batch. The lower limit of quantification was 20 ng/mL. Inter- and intra-day coefficients of variation were below 9% for all quality control samples.® . Given the sparse data (1-3 observations per participant), a model developed using rich stavudine concentration data from a separate study our group has conducted of African adults from the same community was used [3/4 (between subject variability 17%CV), apparent volume of distribution 33.5 L/kg, first-order absorption rate constant 11.1/h (between subject variability 125%CV), absorption lag time 0.41 h, proportional residual variability 27% and additive residual variability 10 ng/mL. Individual pharmacokinetic parameter sets were then obtained using Bayesian estimation given these model parameters and the observed data. The geometric mean ratio (GMR) was calculated by comparing the individual log-transformed AUC to the mean log-transformed AUC of the overall population.The aim of the pharmacokinetic analysis was to obtain a prediction of each participant's apparent stavudine clearance (CL/F) for calculation of the area under the concentration curve (AUC), where AUC = dose (ng)/clearance (L/h). The data were analysed using nonlinear mixed effects modelling with NONMEMwas used . The pop2 test (or Fisher's exact test), and continuous data were compared using student's T-test or Mann-Whitney test, whichever was appropriate. Logistic regression models were fitted to assess the association between GMR > 1 and IFG, IGT, hyperlactataemia, hypertriglyceridaemia and lipoatrophy. Linear regression models were fitted to assess the association between log-transformed stavudine area under the curve and the following variables: concentrations of glucose, lactate and triglycerides, and lipoatrophy scores. All tests were two-sided, and a P-value < 0.05 was considered significant. Analyses were performed using SPSS Means (standard deviation (S.D)) and medians (interquartile range) were used to describe parametric data and non-parametric data, respectively. Categorical data were compared using χ2, respectively. Median waist to hip ratio was 0.85 (0.80 to 0.92). Median (IQR) current CD4 count was 304 (234-516) cells/μL. Twelve participants were virologically suppressed, 6 had viral load above 50 copies/mL and 29 had no viral load data. Forty and seven participants were on 30 mg and 40 mg of stavudine, respectively. Twenty six, twenty and one participants were on efavirenz, nevirapine and lopinavir, respectively. All participants were on lamivudine. The median (interquartile range (IQR)) fasting glucose concentration was 4.9 (4.7 to 5.4) mmol/L and the mean (standard deviation (sd)) 2 hour glucose concentration was 5.34 (1.43) mmol/L. Nine and two participants had IFG and IGT, respectively. The mean (sd) lactate concentration was 2.26 (0.78) mmol/L and 15 participants had hyperlactataemia. The median (IQR) triglyceride concentration was 1.17 (0.85 to 1.60) mmol/L and 11 participants had hypertriglyceridaemia. The median (IQR) lipoatrophy score was 0 (0 to 9) and 16 patients had lipoatrophy.Forty seven black participants were included for the analysis. Median (IQR) age was 34 (30-38) years. Thirty nine participants were female. Median (IQR) weight and body mass index were, 61.0 (54.4 to 73.8) kg and 24.5 (21.5 to 30.4) kg/mA total of 122 stavudine concentrations from 47 participants were analysed. Eleven participants had no pre-dose concentrations because they took their stavudine morning doses prior to the OGTT. The observed stavudine concentrations plotted against the model predictions are shown in Figure We found no association between log-transformed stavudine AUC and metabolic parameters expressed as continuous variables Table . We alsoDespite guidelines recommending that the use of stavudine be avoided because of its toxicity, it continues to play a critical role in scaling up antiretroviral therapy in resource poor settings. Therefore, studies examining pathogenesis of stavudine toxicity are still relevant. We found no association between stavudine AUC and the lipoatrophy scores or concentrations of glucose, lactate and triglycerides. To our knowledge this is the first study to evaluate the association between plasma stavudine concentrations and serum glucose, lactate and triglycerides. We found high prevalence of metabolic abnormalities in this black African cohort with a median duration of stavudine exposure of 14.5 months: lipoatrophy (34%), dysglycaemia (23%), hyperlactataemia (32%), and hypertriglyceridaemia (23%). We found an association between duration and triglycerides concentrations.A meta-analysis from randomised control trials and cohort studies showed that switching from higher to lower doses of stavudine, or starting at lower doses, is associated with improvement in stavudine toxicity without loss of efficacy . SwitchiAlthough stavudine related toxicity is well documented, to date, few studies have investigated pharmacokinetic relationship with stavudine toxicity. Our findings are different to a case-control study conducted by ter Hofstede et al, which reported that cases with lipoatrophy had higher stavudine exposure than controls . HoweverOur study had a few limitations. First, we measured stavudine concentrations in plasma and not the active intracellular triphosphorylated metabolite. Second, we used sparse sampling instead of intensive sampling. However, a population approach allowed us to predict individual AUCs, an acceptable measure of drug exposure. Third, we did not have data on genetic polymorphisms. Fourth, sample size of this study was small, and therefore might have insufficient power to detect relatively small effects of plasma concentrations on metabolic abnormalities. However, this sample size is larger than in other pharmacokinetic studies that have examined the association between stavudine concentrations and metabolic toxicity ,22.Future studies examining the pathogenesis of stavudine-associated toxicities should have adequate power and preferably be longitudinal. Relevant genetic studies should also be done in the populations where stavudine will still be used in the medium term. Physiologically based pharmacokinetic models that take into account the temporal fluctuations and intracellular cascade steps of plasma NRTIs and metabolites should be used to establish pharmacokinetic-pharmacodynamic relationships.In conclusion, we did not find an association between stavudine exposure and metabolic complications. Despite guidelines recommending that the use of stavudine be avoided because of its toxicity, it is still widely used in resource poor settings. Until there is universal access to safer drugs, there is a need for further studies examining the pathogenesis of stavudine-associated toxicities.The authors declare that they have no competing interests.PZS participated in the study design, acquisition of data, data analysis and interpretation, and drafted the manuscript. JSvdW participated in study design, population pharmacokinetic analysis and helped to draft and critically revise manuscript. HMM participated in study design, data interpretation, and critical revision of the manuscript. MB performed statistical analysis and helped to draft and revise manuscript. PJS performed analysis of the samples and helped to draft the manuscript. JAD participated in study design, acquisition of data and critically revised the manuscript. NSL participated in study design and acquisition of data. GM conceived of the study, participated in study design, data interpretation, and critically revised manuscript. All authors read and approved the final manuscript.


A total of 474 subjects who were free of liver cancer and cirrhosis and were initially selected as controls for previous case–control studies of aflatoxin-induced hepatocarcinogenesis in Taiwan, were employed in this study. Aflatoxin-albumin adducts were determined by competitive enzyme-linked immunosorbent assay, hepatitis B surface antigen and antibodies to hepatitis C virus by enzyme immunoassay, as well as genotypes of glutathione S-transferase M1-1 and T1-1 by polymerase chain reaction. The detection rate of AFB1-albumin adducts was significantly higher in males (42.5%) than in females (21.6%) . The formation of detectable albumin adducts was moderately higher in hepatitis B surface antigen carriers (42.8%) than in non-carriers (36.6%) . In addition, the detection rate of AFB1-albumin adducts tended to increase with the increasing number of null genotypes of glutathione S-transferase M1-1 and glutathione S-transferase T1-1. In conclusion, this cross-sectional study has assessed the relative contributions of environmental exposure and host susceptibility factors in the formation of AFB1-albumin adducts in a well characterised Chinese adult population. This study further emphasises the necessity to reduce aflatoxin exposure in people living in an area endemic for chronic hepatitis B virus infection.Dietary exposure to aflatoxins is one of the major risk factors for hepatocellular carcinoma. Individual susceptibility to aflatoxin-induced hepatocarcinogenesis may be modulated by both genetic and environmental factors affecting metabolism. A cross-sectional study was performed to evaluate determinants of the formation of aflatoxin covalently bound to albumin  87, 966–970. doi:10.1038/sj.bjc.6600584www.bjcancer.comCancer Research UK© 2002  Although chronic infection with hepatitis B virus (HBV) is now regarded as the major cause of HCC in high-incidence areas , or AFP (20 ng ml−1), or a family history of HCC or liver cirrhosis among first degree relatives was referred for the second-stage screening by upper abdominal ultrasonography. The abdominal ultrasonography was performed by board-certified gastroenterologists who were well experienced in ultrasonographic examinations using Toshiba SAL-38B and SSA-240A ultrasonographic apparatus with 3.75 MHZ real-time linear and sector probes .From July 1990 through June 1992, a community-based two-stage liver cancer screening programme was carried out in seven townships in Taiwan. The cohort characteristics and methods of screening and follow-up have been described previously  when assaying the equivalent of 200 μg albumin per well was 0.01 fmol μg-1. Samples were assayed by duplicate analysis in duplicate wells. Samples with <20% inhibition were considered non-detectable. Two control samples were analysed with each batch of sera, a pooled sample of plasma from non-smoking US subjects and a positive control of serum from a rat treated with 1.5 mg AFB1.An enzyme-linked immunosorbent assay was used to determine the level of AFBGSTM1-1 genotyping for gene deletion was performed by PCR amplification with primers for exons 6 and 7, which produced a 210 bp band, according to the method of GSTT1-1 genotype was determined using the technique of 1-albumin adducts and various variables. In addition, months of year for blood sample collection were grouped into four seasons in order to evaluate seasonal variations in the detectable levels of AFB1-albumin adducts. All analyses were performed with SAS software  and all P values for tests of statistical significance were based on two-tailed probability.Because it was not considered appropriate to assign a value to the undetectable serum level of AFB1-albumin adducts, the adducts level was analysed as a binary rather than continuous variable. Odds ratios (OR) and their 95% confidence intervals (CI), which were derived from logistic regression models, were used to indicate the magnitude of the associations between formation of AFB1-albumin adducts with these demographic characteristics are described in Table 11-albumin adducts among study townships (ranging from 33.3 to 47.5%), seasonality of sample collection (ranging from 36.8 to 43.0%) and age groups (ranging from 26.9 to 40.6%). In contrast, males had significantly higher detection rate of AFB1-albumin adducts than females (42.5 vs 21.6%) with an odds ratio (OR) of 2.7 (95% CI=1.5–5.0).The demographic data concerning the study subjects and the relationship of the positivity of AFB1-albumin adducts in relation to multiple HCC risk factors are summarised in Table 21-albumin adducts were detectable in 42.8% (86 of 201) of HBsAg carriers and 36.6% (100 of 273) of HBsAg non-carriers. The difference in the detection rate was moderately significant between the two groups . This detection rate was higher in study subjects who smoked cigarettes (43.8%) than in those who never smoked cigarettes (35.1%), with a moderately significant OR of 1.4 (95% CI=1.0–2.1). In addition, there were also moderately significant differences in the adduct detection rate depending on GSTM1-1 and GSTT1-1 genotypes; the detection rate was higher in individuals with either GSTM1-1 null (43.0%) or GSTT1-1 null (44.2%) genotype than in those with GSTM1-1 (36.4%) or GSTT1-1 (35.9%) present. The OR of detectable AFB1-albumin adducts associated with GSTM1-1 or GSTT1-1 null genotype was 1.3 (95% CI=1.0–1.9) and 1.4 (95% CI=1.0–2.0), respectively. On the other hand, the detection rate was lower in anti-HCV-positive subjects (30.6%) than negative subjects (41.6%) . There was again a slightly lower detection rate in individuals who consumed alcohol (36.2%) than in those who never drank alcohol (39.9%).Results with regard to the detection rate of AFBGSTM1-1 and GSTT1-1 genotypes on the detectable adduct levels was then analysed and stratified by HBsAg status. In this case the positivity of AFB1-albumin adducts tended to increase with increasing number of the null genotype of GSTM1-1 and GSTT1-1 in both the HBsAg carrier and non-carrier groups, albeit the trend was statistically non-significant . Whereas, no significant association with detectable AFB1-albumin adduct levels was observed for age, anti-HCV positive status, habits of cigarette smoking and alcohol intake, and the null genotype of GSTM1-1 and GSTT1-1.Results of logistic regression analysis of multiple factors associated with the positivity of AFB1 is one of the major risk factors in the multifactorial etiology of HCC . In particular, HBsAg-positive males have 50% higher frequency of detectable adducts than their HBsAg-negative counterpart . The possible association between chronic HBV infection and the increased activation of AFB1 has been examined in epidemiological studies. Two studies of young Gambian children have reported higher AFB1-albumin adduct levels in HBsAg carriers than in non-carriers  and GST-θ (GSTT1-1), are important candidates for involvement in susceptibility to aflatoxin-related liver cancer because they may regulate an individual's ability to metabolize the ultimate carcinogen of aflatoxins, the exo-epoxide (1-albumin adducts tended to increase as the number of null genotypes of GSTM1-1 and GSTT1-1 increased. This biological gradient was observed in both HBsAg carriers and non-carriers, albeit the trend was statistically non-significant.Accumulating evidence indicates that genetic polymorphisms in AFB1 macromolecular adducts in humans is still limited. In this study, we have assessed the relative contributions of environmental determinants and host susceptibility factors in the formation of aflatoxin–albumin adducts in a well characterised Chinese adult population. The result of present study suggests that gender and HBsAg carrier status are major determinants of the formation of aflatoxin covalently bound to albumin. This study further emphasises the necessity to reduce aflatoxin exposure in people living in an area endemic for chronic HBV infection.In essence, our knowledge base about determinants of formation of AFB


Stylophora pistillata were lopped of three different genets, their skeletons marked with alizarin red-S, and divided haphazardly into three morphometric treatment groups: (I) upright position; (II) horizontal position, intact tip; and (III) horizontal position, cut tip. After 1 y of in-situ growth, the 45 surviving ramets were brought to the laboratory, their tissues removed and their architectures analyzed by 22 morphological parameters (MPs). We found that within 1 y, isolated branches developed into small coral colonies by growing new branches from all branch termini, in all directions. No architectural dissimilarity was assigned among the three studied genets of treatment I colonies. However, a major architectural disparity between treatment I colonies and colonies of treatments II and III was documented as the development of mirror structures from both sides of treatments II and III settings as compared to tip-borne architectures in treatment I colonies. We did not observe apical dominance since fragments grew equally from all branch sides without documented dominant polarity along branch axis. In treatment II colonies, no MP for new branches originating either from tips or from branch bases differed significantly. In treatment III colonies, growth from the cut tip areas was significantly lower compared to the base, again, suggesting lack of apical dominance in this species. Changes in branch polarity revealed genet associated plasticity, which in one of the studied genets, led to enhanced growth. Different genets exhibited canalization flexibility of growth patterns towards either lateral growth, or branch axis extension . This study revealed that colony astogeny in S. pistillata is a regulated process expressed through programmed events and not directly related to simple energy trade-off principles or to environmental conditions, and that branch polarity and apical dominance do not dictate colony astogeny. Therefore, plasticity and astogenic disparities encompass a diversity of genetic (fixed and flexible) induced responses.The high morphological resemblance between branching corals and trees, can lead to comparative studies on pattern formation traits, best exemplified in plants and in some cnidarians. Here, 81 branches of similar size of the hermatypic coral  In multicellular organisms, the level of integration among bodily components dictates the final functional performance of the entire organism A further challenging topic is the study of organisms' architectures, made of multiple genetically identical modules, at several levels of organization, which are physiologically and structurally integrated Pseudopterogorgia bipinnata, Sànchez and Lasker In various modular organisms, including plants In branching corals, architectural characteristics can be deduced from traits at three hierarchical levels of organization, the individual polyps Stylophora pistillataS. pistillata, characterized by a continuum of architectural design with several distinct stages. Each stage was marked by its own characteristic morphometric parameters. We presupposed S. pistillata. Special attention is given to whether apical dominance and branch orientation are important in ruling colonial architectures of branching-corals'.To elucidate further the rules that govern colony development in branching forms, we studied plasticity of colony astogeny and branch to colony trajectories in the Indo-Pacific branching coral in situ growth, 45 of the 81 fragments (55.6%) survived and developed into colonies of various shapes. These included 15 colonies from treatment I , 14 colonies from treatment II  and 16 colonies from treatment III . Upon collection, each colony was dried, measured, and photographed from all angles and 22 morphometric parameters (MPs) were taken , treatment II (42±14%) and treatment III (23±3%) differed significantly from each other  as compare to genotype H (6.01±1.04gr) and J  between tips or bases of the two polarity manipulated treatments (II compared to III); and (2) between tip (either intact or cut-off) and base of the same branch. Five most relevant MPs for this comparisons were chosen: (1) nB ; (2) %nB (percentage of branches developed from tip (TBB) or from base (BBB) of the total number of branches in the colony); (3) TBL ; (4) EV ; and (5) %Nx, the number of branches from orders 2 to 4 as part of the total number of branches .Comparisons between the treatments revealed no significant differences in any tested MPs between treatments II and III for all the genets.3) compared to the base  revealed no significant differences between all five MPs studied in genets H and I  to coral colony astogeny and to address the possible existence of apical dominance, a well-documented phenomenon in the plant world, in shaping coral architectures. Results of this, and earlier studies on the branching coral Stylophora pistillata ramets, grown from similar size branches, and sub-cloned from three different coral genets. Ramets were haphazardly divided into three morphometric settings  or treatment II colonies , genotype-based differences emerged in treatments III . We found that altering branch orientation, in addition of trimming the branch tip (upper 0.5 cm), triggered species-specific and colony-specific architectural reactions. Genotype H exhibited an increase in percentage of down-facing branches (%DGB) as compared to the other two genotypes, genotype I gained more weight (W1) as compared to the other genotypes and genotype J showed an increase in percentage of branch order 2 (%N2) as compared to the other two genotypes. Other MPs, where differences were not yet resulted in significant values (because of high variation in the results) and were noticeable to the eye, were the developed total number of branches (nB) and ecological volume (EV). Genotype H showed 19% and 15% increase in total number of branches when comparing treatment I to II and I to III, respectively, and 45% and 7% in ecological volume, respectively. Genotype I exhibited increases of 115% and 143% in total number of branches when comparing treatment I to II and I to III, respectively, and 123% and 186% in ecological volume, respectively. Genotype J displayed 44% and 17% increase in nB, respectively, 27 and 32% and 39% in EV, respectively. As stated, these values however, were not significantly different from each other's.S. pistillata mode of astogeny S. pistillata branching system, as in the case of the soft coral Nephthea sp., where cutting the terminal polyps in young colonies did not change the way the colony developed While the traditional test for apical dominance in plants entails removal of the apical bud and measuring the effects on the dormant buds lower on the branch Stylophora settings II and III in the present study, Acropora branches developed terminal polyps at both sides of the branch. However, while in the Acropora treatment Stylophora treatment did not show any phototropism impact as new branches initiated up and down trajectories, without obvious preference for light.The impacts of branch orientation and polarity on morphometric parameters regulating colonial astogeny was revealed earlier by Kawaguti Stylophora species-specific colonial astogeny from vertical oriented branches, most new branch initiations developed along the branch.Changing branch orientation affected the way morphology developed, the complexity of the branching system (represented by numbers of branch generations), and the growth of new branches from branch tip or along the branch. When the initial branch orientation changed from horizontal to vertical, the percentage of high order of branches (N2 compare to N3 or N4) was higher than that of the horizontal branch. Altering the orientation of the branch from vertical to horizontal might result in shifting energy (evident by developing of new branches) from the original branch tip area (intact or cut-tip branch) to the opposite branch end, the base. It is of major interest to note that new branches developed from both tip ends rather than from the whole length of the branch, resembling induced positional information Stylophora pistillata is highly regulated as in whole colony scenarios Acropora millepora and Pocillopora damicornisAstogeny of cut branch in Stylophora pistillata, Meroz et al. It is evident that coral colonies, isolated branches, and spat not only ‘sense’ environmental cues but also ‘discern’ their special position. Working on two branching coral species, including Stylophora pistillata , were carefully detached from the substrates by chisel and hammer. Each colony represented only a single genet, as colonial fragments of this species in Eilat do not resume development in situ in clear plastic bags with alizarin Red S solution  sinTwenty-two morphometric parameters MPs;  were meaFirst step in the analysis carried out “Pearson Correlation”, in order to remove parameters that are related (p>0.9), therefore do not add information to the analysis. The second step of the analysis was to elucidate those MPs that provide the best discrimination between the groups, using Discriminant Analysis test. This analysis was preformed initially within each genotype between treatments and than, on each treatment between genotypes and for each test, three to four most discriminating MPs were chosen. Following that, an ANOVA was performed on the selected MPs checking for significance of differences, first in each genotype between treatments and than, in each treatment between genotypes. The level of significant was calculated using Bonferoni correction to avoid type I error. In the ANOVA tests, the preliminary assumption was the existence of homogeneity of variance and not normal distribution


The summary-estimate for dizygotic twins was 1.3 (1.0–1.7) and for monozygotic or same sex twins 1.4 (1.2–1.8).In a meta-analysis of testicular cancer in twins, twins had a 30% increased risk , providing indirect support for the hypothesis that  The relative importance of environmental factors and genetic factors are not clear, but it is likely that both play an aetiological role  using combinations of individual search terms and subject headings. We inspected the reference lists of studies retrieved from these initial searches for previously unidentified studies, which were then located on Medline to identify their subject headings. Further searches were then undertaken using these derived subject headings.We included all studies that had reported the risk of testicular cancer in twins compared with the general population or a specific single birth control group. Some studies did not provide a risk estimate for all male twins, instead reporting monozygotic and dizygotic twin risks separately , which tests that the true pooled effect is zero. Tests for heterogeneity were based on the We identified seven studies, which met our inclusion criteria . The pooled estimate of risk was 1.31 (1.1–1.6) assuming random effects and 1.34 (95% CI 1.2–1.5) assuming fixed effects. Excluding the earlier Swedish study resulted in an estimate of 1.3 (1.0–1.7) (P=0.068). When we additionally excluded the study with only four cases, we found a risk of testicular cancer among twins of 1.4 (1.2–1.6). We conducted a further analysis excluding the largest and smallest studies, generating a risk estimate of 1.4 (1.1–1.7). Removing the only case–control study did not materially alter the summary estimate. There was no evidence of publication bias .Six of the seven studies reported an increased risk of testicular cancer among twins , with twFive studies either reported the dizygotic twin risk or the risks for male twins that came from an opposite sex twin pair. Pooling these risks resulted in a similar estimate of risks as for all twins . The summary estimate for monozygotic or same sex twins was 1.4 (1.2–1.8).Only two studies have previously found a significant positive association between twinning and the development of testicular cancer, with four others generating suggestive but non-significant results and one finding a non-significant protective effect. These analyses indicate that these are unlikely to be chance findings, that the result is not sensitive to the inclusion/exclusion of eligible studies, and that there is a real increase in testicular cancer among twins.in utero oestrogen exposure may be the underlying cause  that may alter risk of testicular cancer. These include different levels of many other hormones and characteristics of the mother including age, parity, and height . Twins oThere is emerging evidence that being a twin is protective for childhood cancer, possibly due to the lower birth weight of twins, earlier limitation of their growth velocity, or the selection effects of higher intrauterine death rates . Hence a