Attitudes towards intimate partner violence against women (IPVAW) has been suggested as one of the prominent predictor of IPVAW. In this study, we take a step back from individual-level variables and examine relationship between societal-level measures and sex differences in attitudes towards IPVAW.We used meta-analytic procedure to synthesize the results of most recent data sets available from Demographic and Health Survey (DHS) of 17 countries in sub-Saharan Africa conducted between 2003 and 2007. Pooled odds ratio (OR) and 95% confidence intervals (CI) were computed for all countries. Test of heterogeneity, sensitivity analysis, and meta-regression were also carried out.Women were twice as likely to justify wife beating than men  with statistically significant heterogeneity. The magnitude in sex disparities in attitudes towards IPVAW increased with increasing percentage of men practicing polygamy in each country. Furthermore, magnitude in sex disparities in attitudes towards IPVAW decreased monotonically with increasing adult male and female literacy rate, gender development index, gross domestic product and human development index.This meta-analysis has provided evidence that women were more likely to justify IPVAW than men in sub-Saharan Africa. Our results revealed that country's socio-economic factors may be associated with sex differential in attitudes towards IPVAW. Intimate partner violence against women (IPVAW) is present in almost all societies ,2 and isFew studies have examined gender differences in attitudes towards IPVAW in sub-Saharan Africa (SSA) ,18,19. TThis meta-analysis used data from Demographic and Health Surveys (DHS) conducted between 2003 and 2007 in sub-Saharan Africa available as of November 2008. DHS surveys were implemented by respective national institutions and ICF Macro International Inc. with financial support from the US Agency for International Development (USAID). Methods and data collection procedures have been published elsewhere . BrieflyTo assess the degree of acceptance of wife-beating by women and men, respondents were asked the following question: "Sometimes a husband is annoyed or angered by things which his wife does. In your opinion, is a husband justified in hitting or beating his wife in the following situations?" The five scenarios presented to the respondents for their opinions were: "if wife burns the food," "if wife argues with the husband," "if wife goes out without informing the husband," "if wife neglects the children," and "if the wife refuses to have sexual relations with the husband". Information was collected from all women and men irrespective of their marital status. A binary outcome variable for acceptance of wife-beating was created and coded as '0' if the respondent did not agree with any of the above mentioned five scenarios or did not have any opinion on the issue and coded as '1' if the respondent agreed with at least one scenario.We gathered country-level data matched within the same time frame when DHS were conducted from the reports published by the United Nations Development Programs and WorlI2 describes the percentage variation across studies that have heterogeneity [I2 were adjusted to zero (no heterogeneity) to give an I2 between 0 and 100%, where larger values show increasing heterogeneity. We performed leave-one-country-out sensitivity analysis to determine the stability of the results. This analysis evaluated the influence of individual studies by estimating the weighted average OR in the absence of each country.We calculated Odds ratios (OR) for the association between sex of the respondent and acceptance of IPVAW for each country. We used DerSimonian-Laird method (random-effects model) to calcuogeneity ,32. Negap < 0.05 was considered significant. Stata 10  software was used for analysis.We investigated the impact of various country characteristics on OR estimates through an inverse-weighted linear meta-regression analysis. The independent variable was the natural logarithm of the OR. The explanatory factors included the country characteristics (listed above), sample size, sub-region and the calendar year of the survey. This analysis accounted for aspects such as effect modifications of the explanatory factors by performing univariable linear regression analyses for each factor. All tests were two sided and This study was based on an analysis of existing survey data with all identifier information removed. The survey was approved by the Ethics Committee of the ICF Macro at Calverton in the USA and by the National Ethics Committees in their respective countries. All study participants gave informed consent before participation and all information was collected confidentially.p = 0.001) and the corresponding I2 (99%) indicated statistically significant heterogeneity. In the leave-one-country-out sensitivity analysis the CIs did not change materially with exclusion of any of the countries, which remains within 95% confidence interval of the overall estimate for all countries  limitations. In meta-analysis, the traditional unit of analysis is each study (country in our case), thus, compared with multilevel analysis with individual-level data, the power to detect a difference in aggregate or to identify explanatory variables by meta-regression is greatly diminished. As with all ecological studies, the findings of this study cannot be considered conclusive because of the cross-sectional and ecological design and the possibility of ecological fallacy. Thus, caution should be exercised in the attribution of a casual relationship and the direction of relationship observed in the study. Another limitation is that the meta-regression analysis is based on univariable analysis, due to small sample size (number of countries included), the study could not control for potential confounders simultaneously using multivariable meta-regression analysis. Despite these limitations, the study strengths are significant. It is a large, population-based study with national coverage. It is increasingly recognised that, even when studying individual level risk factors, population level studies play an essential part in defining the most important public health problems to be tackled, and in generating hypotheses as to their potential causes. An impoWe directly analysed substantial number of public domain data sets instead of using results from published studies. This approach, we believe, brings with it two considerable advantages. First, not all published studies include the same or even comparable variables in their analyses. Taking raw data from DHS allow us to use the same variables and most comparable items for attitudes toward IPVAW. Second, there are many more survey data sets than there are published studies. We therefore, achieve a much greater coverage of the population of effect sizes and mitigate the 'publication bias' that increases the probability of Type I errors. This isThis meta-analysis has provided evidence that women were more likely to justify IPVAW than men in sub-Saharan Africa. We found that the better the country economic status, adult female and male literacy rate, gender development index, human development index the higher the sex disparities in attitudes toward IPVAW. However, it is important to note that there was high heterogeneity in sex differences in attitudes towards IPVAW across countries. Thus, suggesting that multifaceted geographically differentiated intervention may represent a potentially effective approach for addressing issues related to IPVAW in sub-Saharan Africa and policies have to be tailored to country-specific conditions.The authors declare that they have no competing interests.OAU, SL and TM were involved in the conception of the study. OAU carried out data extraction. OAU conducted statistical analysis under supervision of SL and TM. OAU drafted the paper with contributions from the co-authors. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2458/10/223/prepub


Since 2002, active surveillance programmes have detected numerous atypical scrapie (AS) and classical scrapie cases (CS) in French sheep with almost all the PrP genotypes. The aim of this study was 1) to quantify the genetic risk of AS in French sheep and to compare it with the risk of CS, 2) to quantify the risk of AS associated with the increase of the ARR allele frequency as a result of the current genetic breeding programme against CS.We obtained genotypes at codons 136, 141, 154 and 171 of the PRNP gene for representative samples of 248 AS and 245 CS cases. We used a random sample of 3,317 scrapie negative animals genotyped at codons 136, 154 and 171 and we made inferences on the position 141 by multiple imputations, using external data. To estimate the risk associated with PrP genotypes, we fitted multivariate logistic regression models and we estimated the prevalence of AS for the different genotypes. Then, we used the risk of AS estimated for the ALRR-ALRR genotype to analyse the risk of detecting an AS case in a flock homogenous for this genotype.Genotypes most at risk for AS were those including an AFRQ or ALHQ allele while genotypes including a VLRQ allele were less commonly associated with AS. Compared to ALRQ-ALRQ, the ALRR-ALRR genotype was significantly at risk for AS and was very significantly protective for CS. The prevalence of AS among ALRR-ALRR animals was 0.6‰ and was not different from the prevalence in the general population.In conclusion, further selection of ALRR-ALRR animals will not result in an overall increase of AS prevalence in the French sheep population although this genotype is clearly susceptible to AS. However the probability of detecting AS cases in flocks participating in genetic breeding programme against CS should be considered. Bovine spongiform encephalopathy (BSE) is a zoonotic transmissible spongiform encephalopathy (TSE), which may have spread to sheep and goat populations. This situation has prompted European countries to organise control and surveillance of TSEs in small ruminants in which BSE cannot be clinically distinguished from scrapie.C) [In sheep, genetic susceptibility to classical scrapie (CS) is supported by polymorphic variations at codons 136, 154 and 171 of the PRNP gene, which encodes the normal cellular protein prion (PrPC) . This knC) ,3.Recently, an atypical form of scrapie (AS) has been detected and has challenged such programmes since animals genetically resistant to CS are affected -7. HowevFurthermore, most AS cases have been observed in animals with genotypes showing low susceptibility to CS. Several studies have shown that AS susceptibility is highly associated with PrP codons 141 (L/F) and 154 (R/H) ,18. It iA descriptive study has reported that the estimation of AS prevalence in Europe is similar, whatever the country or tested population and amounts to about 6 cases per 10,000 tests. Conversely, CS prevalence appears to be heterogeneous both between countries and, within countries, between fallen stock and healthy slaughter . This coWe have carried out a case control study by using data from the TSEs surveillance programme to estimate the risk of AS associated with the PrP genotypes in the French sheep population and to compare it with the risk of CS. In addition, we have used our dataset to estimate the prevalence of AS per genotype and the risk for a fully CS resistant flock (100% ALRR-ALRR) to be detected positive for AS under certain assumptions.st January 2002 to 31st December 2007. Each year since 2002, a number of adult (over 18 months old) sheep from unknown TSE status flocks was randomly selected for TSE detection at abattoirs  or at rendering plants . Details concerning the French active surveillance programme are described in Cazeau [Cases were recruited among the animals tested for TSEs in France from 1n Cazeau . Betweenn Cazeau .All positive samples were confirmed and typed by the National reference laboratory for animal TSEs of AFSSA Lyon using EU approved reference methods . For all17025). PrP polymorphisms at codons 136, 141, 154 and 171 of the 248 AS cases were analysed after direct sequencing of the coding region of the PRP gene encompassing codons 92-282. These analyses were conducted by INRA . For more details concerning primers and sequencing see [During that period, the PrP genotype of all cases was routinely determined, when suitable samples were available, at codons 136, 154 and 171 with a technique that could not differentiate ARH and ARQ alleles. Further analyses were conducted to specify alleles at position 141 and 171 by the Taq man method ,24. It icing see .Controls were recruited among animals that were tested for the presence of TSEs with a negative result. Controls for AS were animals tested with one of the tests recommended for the detection of AS and controls for CS were animals tested with any rapid test.As a legal requirement , during In order to adjust for potential confounding variables, we collected information on the stream of surveillance  and the dentition of the animals (2-4 definitive incisors (DI)/5-7 DI/8 DI) which is a poor proxy for age  but was For each type of scrapie, we fitted multivariate logistic regression models to assess the risk associated to genotypes with ALRQ-ALRQ as the baseline category, because this genotype is assumed to be the ancestral genotype. We adjusted the models on the surveillance stream with healthy slaughter as the baseline category, and on dentition, with the class corresponding to the youngest animals as the baseline category. The model outputs were odds ratio (OR) which can be interpreted as an approximation of the relative risk in the context of rare diseases. We could not estimate OR associated with the genotypes in which no case was detected. The value of the OR and their 95% confidence intervals for the 1,000 datasets were computed using the Rubin method .We estimated the prevalence of AS for each genotype. Because the proportion of animals genotyped was different for cases and controls, we defined it as the national prevalence  times the proportion of the genotype among cases, divided by the frequency of the genotype in the general population. Using a Bayesian approach, we considered that each of these proportions followed a beta distribution with mean and variance estimated by the observations. We derived the median and the 95% credibility interval of prevalences based on 2,000 iterations obtained after convergence.n animals tested was:We have considered the risk for a fully ALRR-ALRR flock to be detected positive for AS. We assumed that in any flock the individual risk of AS depended only on the genotype, the age and the surveillance stream. Considering that the animals tested from a given flock had the same distribution of age and surveillance stream as the general population, the average individual risk for an ALRR-ALRR animal could be estimated by the prevalence previously estimated for this genotype. Since the occurrence of AS in a flock was defined as an independent event, the number of cases in a flock, X, followed a binomial distribution. The probability for at least one animal being AS positive among Using the previous methods to estimate the prevalence per genotype, we derived the median of the probability and its 95% credibility interval for n varying from 1 to 1,000, based on 2,000 iterations obtained after convergence.All the statistical analyses were done with R for Windows  and WinbAfter exclusion of missing information on genotypes and dentition, 248 AS cases, 245 CS cases and 3,317 controls were used to estimate OR and prevalence  were respectively 11, 25 and 31 times higher than the prevalence for the general population , the oldest animals were more at risk but OR was not significantly different from 1 at 5%. However, this could be due to a lack of power since comparison with the intermediate category showed that the OR associated with the oldest category was significantly different from 1. For CS there was no observed effect of age.The probability for a 100% ALRR-ALRR flock to be detected as positive for AS was not negligible when 100 animals were tested  was not significantly different from the current average national prevalence . Therefore in standing conditions and assuming AS occurrence is not dependent on infectious exposure, increasing frequency of the ALRR allele in the general population should not result in an increase of the global AS prevalence in the sheep population.Despite the low prevalence of AS in ALRR-ALRR animals, the probability of detecting a positive case is not negligible when the number of animals tested increases as shown by Figure The assumption of a common risk of AS in farms is supported by the homogenous distribution of the disease and similarities of prevalence per genotype in France and in Europe ,19,33. TFinally, AS occurrence appears to be heavily dependent on the PrP genotype, possibly modulated by some environmental risk factors which are to be further explored. Even if the open reading frame of the PRNP gene has a strong influence on the occurrence of AS it might not be the unique genetic factor of susceptibility. Other regions of the PRNP or other genes could be involved either independently or in synergy and this hypothesis is also worth to be investigated.The authors declare that they have no competing interests.AF, PG, CD, DC and CM conceived the study, participated in its design and coordination. AF and CM performed the strategy of the selection of animals and carried out the statistical analysis. JNA performed the scrapie status confirmation and the strain determination. PL and KMG performed the polymorphism genotyping work by sequencing. AF, PG, CD, DC, CM, JNA and KMG helped to draft the manuscript. All authors read and approved the final manuscript.


APOE genotypes, cardiovascular risk factors, and a non-invasive measure of arterial stiffness in the Brazilian population.Cardiovascular diseases (CVD) are the main cause of death and disability in developed countries. In most cases, the progress of CVD is influenced by environmental factors and multifactorial inheritance. The purpose of this study was to investigate the association between APOE polymorphism was conducted by PCR-RFLP and pulse wave velocity analyzed with a noninvasive automatic device.A total of 1493 urban Brazilian individuals were randomly selected from the general population of the Vitoria City Metropolitan area. Genetic analysis of the Age, gender, body mass index, triglycerides, creatinine, uric acid, blood glucose, blood pressure phenotypes were no different between ε2, ε3 and ε4 alleles. The ε4 allele was associated with higher total-cholesterol (p < 0.001), LDL-C (p < 0.001), total-cholesterol/HDL-C ratio (p < 0.001), LDL/HDL-C ratio (p < 0.001), lower HDL-C values (p < 0.001) and higher risk to obesity  and hyperuricemia . Nevertheless, pulse wave velocity (p = 0.66) measures were no different between genotypes. The significant association between APOE genotypes and lipid levels persisted after a 5-year follow-up interval, but no interaction between time and genotype was observed for lipids longitudinal behavior.APOE gene is associated with a worse lipid profile in the Brazilian urban population. In our relatively young sample, the observed effect of APOE genotype on lipid levels was not translated into significant effects in arterial wall stiffness.The ε4 allele of the  Cardiovascular diseases (CVD) are the main cause of death and disability in developed countries. In most cases, the progress of CVD is influenced by multifactorial inheritance and environmental factors -4. LipopAPOE gene, located on chromosome 19[APOE polymorphism on plasma lipids have shown that the presence of the ε4 allele is associated with elevations in LDL-C, while the presence of ε2 is associated with decreased levels of LDL-C[Apolipoprotein E (ApoE) is a glycoprotein that plays a fundamental role in the lipid metabolism. ApoE participates in the clearance of chylomicron remmants and very low-density lipoprotein (VLDL) by serving as a ligand for LDL receptors. It is amosome 19. The ε3 mosome 19-12. Many of LDL-C. Moreove of LDL-C althoughAPOE gene and knowledge of its genetic variants, several studies have demonstrated the association between the APOE polymorphisms and chronic conditions, such as Alzheimer's disease[APOE genotype groups with the prevalence of the major CVD risk factors and its possible association with the evolution of the studied phenotypes in a longitudinal study of Brazilian subjects randomly selected from an ethnically mixed urban population.After the discovery of the s disease, age-rels disease, osteopos disease, breast s disease, end-stas disease, atheross disease, diabetes disease, coronars disease and longs disease. Based oAPOE polymorphism and that this worse profile would be translated into significantly different measures of arterial stiffness.Our hypothesis was that individuals carrying the ε4 allele had a worse lipid profile when compared with ε2 and ε3 alleles carriers of the A cross-sectional study of risk factors for cardiovascular diseases was performed in the urban population of the Vitória city, Brazil, using the WHO-MONICA project guidelines. In the This study was approved by the Ethics Committee for Research on Human Subject of the Espírito Santo Federal University, and all subjects gave written informed consent to participate.2[Weight and height were measured according to a standard protocol, with participants wearing light clothing and no shoes. Height was measured in centimeters and weight in kilograms using a calibrated balance. Body mass index (BMI) was calculated and obesity defined as BMI ≥ 30Kg/m2. Individ2. Partici2. SubjectBlood pressure was measured in the sitting position with the use of a standard mercury sphygmomanometer on the left arm after 5 minutes' rest. The first and fifth phases of Korotkoff sounds were used for systolic (SBP) and diastolic pressure (DBP), respectively. The SBP and DBP were calculated from two readings with a minimal interval of 10 minutes. Hypertension was defined as mean SBP ≥ 140 mmHg and/or DBP ≥ 90 mmHg or use of anti-hypertension drugs . Pulse pThe Carotid-femoral pulse wave velocity (PWV) was analyzed with a noninvasive automatic device  by an experienced observer blinded to the clinical characteristics. Briefly, common carotid artery and femoral artery pressure waveforms were recorded noninvasively using a pressure- sensitive transducer . The distance between the recording sites (D) was measured, and PWV was automatically calculated as PWV = D/t, where (t) means pulse transit time. Thirty measurements were repeated over 10 different cardiac cycles, and the mean is used for the final analysis. Because systolic BP has direct influence on PWV, we also adjusted PWV for the mean systolic BP in all groups. The validation of this automatic method and its reproducibility has been previously described.Blood glucose, TG, lipoprotein fractions and uric acid were assayed by standard techniques in 12-h fasting blood samples .This stu®  was performed using the primer sequences 5'-GCCGATGACCTGCAGAAG-3' and 5'-CACGCGGCCCTGTTCCAC-3' (fragment size 117 pairs base) and 5'-GCGGACATGGAGGACGTG-3' and 5'- AGCTCCTCGGTGCTCTGG-3' (fragment size 83 pairs base), for rs7412 and rs429358, respectively. In the HRM phase, Rotor Gene 6000®  measured the fluorescence in each 0.1°C temperature increase in the range of 70-94°C. Melting curves were generated by the decrease in fluorescence with the increase in the temperature; nucleotide changes resulting from different curve patterns were analyzed and genotyped  to confirm the genotypes indicated by HRM.Genomic DNA was extracted from leukocytes in samples of whole blood, following a standard salting-out technique. GenotypAPOE polymorphisms (rs7412 and rs429358) was assessed by randomly selecting 96 samples to be re-genotyped using a high resolution melting method  by two independent technicians and these results were 100% concordant.Quality control for the APOE genotypes on cardiovascular risk factors, subjects were categorized into three groups: ε2 carriers (ε2/ε2 + ε2/ε3 genotypes), ε3 carriers (ε3/ε3 genotype) and ε4 carriers (ε4/ε4 + ε3/ε4 genotypes)[To evaluate the effect of enotypes). In eachenotypes).APOE groups were tested by One-Way ANOVA for continuous variables, and χ2 test for categorical parameters. Table Differences in the baseline cardiovascular risk characteristics across the Logistic regression analysis was carried out to estimate the odds ratio (OR), with 95% confidence intervals (CI), in order to assess genetic risk factors for common cardiovascular risk. The risk analysis was performed comparing all three groups together. In the Table Quantitative variables were expressed as the mean ± standard deviation, while qualitative variables were expressed as percentage. Hardy-Weinberg equilibrium for the distribution of the genotype groups was estimated using the Haploview software. All statistical analyses were carried out using SPSS software (v. 16.0), with the level of significance set at p < 0.05.APOE genotypes among 1461 participants were: ε2/ε2 - 0.4%, ε2/ε3 - 9.7%, ε2/ε4 - 2.3%, ε3/ε3 - 61.0%, ε3/ε4 - 24.4% and ε4/ε4 - 2.2%. The allele frequencies were: ε2 - 10.1%, ε3 - 61.0% and ε4 - 26.6%. The genotype distributions for APOE polymorphisms (rs7412 and rs429358) were in Hardy-Weinberg equilibrium (p > 0.05). The ethnicity proportion was different among the APOE allele groups (p = 0.002), where the number of Caucasians is higher than Afro-descendent in the ε3 allele group . Comparative analysis failed to indicate a significant difference among APOE groups and age (p = 0.28), gender (p = 0.13), diabetes (p = 0.81) and smoking status (p = 0.66) , DBP (p = 0.34), MBP p = 0.21), PP (p = 0.34) and PWV (p = 0.66) showed no association with 5, DBP , LDL-C (p < 0.001), TC/HDL ratio (p < 0.001) and LDL/HDL-C ratio were higher in ε4 and ε3 when compared to ε2 allele carriers. However, for TC, LDL-C and TC/HDL-C no difference was observed between ε2 and ε3 alleles (p > 0.05). The HDL-C values (p < 0.001) were lower in ε4 and ε3 alleles when compared with ε2 allele carriers. In addition, ε3 carriers showed higher HDL-C values when compared with ε4 allele carriers. The uric acid values (p = 0.04) were lower in ε2 allele when compared with ε4 allele carriers. Other variables, such as TG (p = 0.28), VLDL-C (p = 0.81), glucose (p = 0.64), creatinine (p = 0.16) and BMI (p = 0.27) failed to show association with APOE groups , low HDL-C , high LDL-C , while ε4 allele confers risk for elevated levels of TC , LDL-C , uric acid  and BMI   are more prevalent in the ε4 allele group than in the ε2 allele group. In another study, Kolovou et al[knock in after a high-fat diet gained more weight than animals that did not express the APOE protein and animals that expressed isoform APOE2. Considering that in vitro receptor binding studies established that lipid-bound ApoE3 and ApoE4 have a similar affinity for LDLr, whereas lipid-bound ApoE2 has a much lower affinity[Classically, obesity and hyperuricemia have shown to have important roles in the development of cardiometabolic disease ,52,53. Tl studies have conns et al. showed tvou et al showed ivou et al. The relvou et al showed t affinity, we hypoAPOE polymorphism and serum uric acid.Our study has potential limitations. First, it should be noted that the relative young age of the studied population may preclude the identification of the association between the studied genotype and measures of arterial stiffness. Second, we did not evaluate the use of cholesterol-lowering drugs, which could confound the observed associations. Finally, if we have measured insulin and urinary uric acid levels in this population, we could shed light into the proposed mechanism for the association of APOE polymorphism and the lipid profile in individuals from the general population, but failed to show this genetic risk factor as an important modulator of arterial distensibility in this same sample.In conclusion, the present study confirms the close association between The authors declare that they have no competing interests.ROA and SRSF participated in the design of the study, performed the statistical analysis and drafted the manuscript. PCJLS and NEF contributed to acquisition of data and its interpretation. RSC, JGM and JEK contributed to conception and design of the study. ACP conceived of the study, participated in its design, coordination and helped to draft the manuscript. All authors read and approved the manuscript.


In meta-analysis, the presence of funnel plot asymmetry is attributed to publication or other small-study effects, which causes larger effects to be observed in the smaller studies. This issue potentially mean inappropriate conclusions are drawn from a meta-analysis. If meta-analysis is to be used to inform decision-making, a reliable way to adjust pooled estimates for potential funnel plot asymmetry is required.A comprehensive simulation study is presented to assess the performance of different adjustment methods including the novel application of several regression-based methods (which are commonly applied to detect publication bias rather than adjust for it) and the popular Trim & Fill algorithm. Meta-analyses with binary outcomes, analysed on the log odds ratio scale, were simulated by considering scenarios with and without i) publication bias and; ii) heterogeneity. Publication bias was induced through two underlying mechanisms assuming the probability of publication depends on i) the study effect size; or ii) the p-value.The performance of all methods tended to worsen as unexplained heterogeneity increased and the number of studies in the meta-analysis decreased. Applying the methods conditional on an initial test for the presence of funnel plot asymmetry generally provided poorer performance than the unconditional use of the adjustment method. Several of the regression based methods consistently outperformed the Trim & Fill estimators.Regression-based adjustments for publication bias and other small study effects are easy to conduct and outperformed more established methods over a wide range of simulation scenarios. Publication bias (PB) has the potential to distort the scientific literature ,2; sinceMore generally, the tendency for smaller studies to show a greater effect than larger studies, when evaluating interventions, has been named small-study effects . These cTypically, the presence of small-study effects in meta-analysis is identified through asymmetry of a funnel plot. A funnel plot is a scatter plot of study effect size  against a measure of study precision on the y-axis   odds ratio scale. The different regression-based methods used to adjust for PB are described below. Additionally, for completeness we also consider the standard fixed and random effects meta-analysis models .Trim & Fill ,13 is pr0 and R0 estimators [0 estimator are presented. We implement this method using fixed effects in both the 'trimming' and 'filling' parts of the algorithm (fixed-fixed), similarly, random effects for both parts (random-random), and fixed effects to 'trim' and random effects to 'fill' (fixed-random). Justification and evaluation of these variants is available elsewhere. [We evaluate both Ltimators  in the ssewhere. ,14et al. [A number of regression models exist that assess the degree of association between the study effect and a measure of its precision. The initial test based on the statistical significance of this association was suggested by Egger et al. , but dueet al. ,16. A fuet al.  is not cEvidence of such an association may suggest that the meta-analysis is affected by PB if the observed smaller, less precise, studies have larger effect sizes than the more precise studies ,18. ThisBy extrapolating this regression line to the point where the standard error is zero, the effect size from a  infinitely large study can be predicted. And, as discussed in Section 1, this can be interpreted as the effect size adjusted for PB. For example in Figure 'Egger' et al. [' et al.  proposed' et al.  can be s' et al. :yi is the ln(OR) from study i (in the context considered in this paper) and sei is its associated standard error. We interpret the two coefficients α and β to represent the adjusted pooled effect (intercept) and the slope associated with funnel plot asymmetry respectively. The regression is weighted by the inverse of the estimated effect size variance for each study (1/sei 2). φ is an unknown multiplicative dispersion parameter that is estimated in the regression and allows for possible heteroscedasticity [where asticity ,24.The notion of this multiplicative error term in this context is not consistent with the typically used variance-weighted meta-regression models where the error is additive . As a reφ in equation (1.a) allows for the incorporation of between-study variability [φ), but, as an additive component of between-study variance (τ2) to the within-study variance (sei 2):This fixed effects regression model assumes homogeneous effect sizes. While the multiplicative dispersion parameter iability , the morμi is a normal error term with mean zero and variance τ2 to be estimated from the data.Here, 'Egger-Var' methods are modifications of the Egger model variants (equations 1.a-c) created by replacing the standard error of each study's effect size with the corresponding variance as the predictor variable. This implies that the relationship between effect size and its variance is linear, whereas the Egger approach assumes linearity in relation to the standard error.'Harbord' is a regression test for small-study effects [ effects . A disti effects . The staZi is the efficient score and Vi is Fisher's information (the variance of Z under the null hypothesis for the ith study).where 'Peters' corresponds to another regression test for funnel asymmetry [et al.'s approach is a modification of Macaskill's test [symmetry . This wel's test  which itl's test , with thl's test .α and β represent the adjusted pooled effect (intercept) and the regression slope respectively. For each study i, a and b represent the observed number who experience the outcome of interest in the treated and control groups, respectively, and c and d are the numbers corresponding to those not developing the outcome in the treated and control group respectively. Thus, the sample size of the ith study corresponds to the sum of ai, bi, ci and di.As before, the two coefficients We suspect, in practice, researchers carry out a test for small-study effects and consider the use of adjustment methods conditional on the outcome of such a test. Therefore, we also evaluate two conditional approaches in which a standard random effects model or either of the original Egger or Harbord adjustment based methods are used depending on whether the corresponding test (i.e. Egger or Harbord respectively) was significant at the 10% level. Since the Egger conditional approach is almost always outperformed by the Harbord conditional method, only the latter is reported below.In summary, the performance of the following adjustment methods is reported below. An abbreviation is given to each method, which is used in the remainder of the paper:• The two usual standard meta-analysis methodsFE)• Fixed effects meta-analysis (RE)•; Random effects meta-analysis • R0 estimator, trim using fixed effects & random effects on filled dataset (TF FE-RE)• R0 estimator, trim using random effects & random effects on filled dataset (TF RE-RE)• R• Parametric adjustment methods: weighted regressions• Egger's model variants:(FE-se);• Fixed effects (RE-se)• Random effects (D-se)• Dispersion • Egger-Var model variants:(FE-var)• Fixed effects (RE-var)• Random effects (D-var)• Dispersion • Other regressions(Harbord)• Harbord's model (Peters)• Peters' model • Conditional methodHarbord-C)• PB test plus conditional adjustment based on it using Harbord's model  and random effects (heterogeneous) meta-analysis scenarios were simulated. Underlying effect sizes  considered were OR = 1 (null effect), 1.5 and 3 (representing a large effect), where OR > 1 is considered clinically beneficial. Following the approach of Schwarzer's et al. , the eveTo simulate each trial within each meta-analysis under fixed effects, the following model was used:δ, and μi is the average event rate on the logit scale in the ith study. ni, piri are the number of subjects, the probability of an event (derived directly from μi and δ), and the number of events in the ith study arm respectively, with a superscript C or T indicating the control or treatment group.Where underlying ln(OR) is Between-study variability (heterogeneity) is simulated here since it has been reported to be an important variable in determining the performance of methods to adjust for PB ,30). Datith trial, δi is assumed to be drawn from a Normal distribution with mean value θ and between-study variance τ2. τ2 is defined to be either 0%, 100%, 150% or 200% of the average within-study variance for studies from the corresponding fixed effects meta-analyses simulated. [Now the underlying effect in the I2 . In a siability .The two most commonly assumed selection processes used in previous simulation studies to induce PB are considered here:The probability that a study is published depending on its resulting p-value is modelled by a step function with discontinuities determined by two cut-points as shown in Table This assumes that only the estimated effect size influences whether a study is included in the meta-analysis or not, so that studies with the most extreme unfavourable estimates of effect are excluded ,14. HereThe meta-analytic scenarios used to investigate the performance of the alternative adjustment methods are intended to encompass a comprehensive collection of realistic situations. Figure Altogether there are 5 different PB situations . Each of these situations is applied to all combinations of the following meta-analysis characteristics: underlying effect size , number of studies in the meta-analysis  and degree of between-study heterogeneity . This results in 240 individual scenarios, with 5,000 simulated datasets generated for each scenario. All methods summarised in section 2.4 were applied to each of these datasets. All statistical analyses were performed using Stata, version 9.2 .Results of this simulation study are reported in section 4; although not all scenarios are shown due to the negligible added value they provide to the overall conclusions. In particular, scenarios with "moderate" bias are omitted because they follow the same overall trend as the more severe scenarios, but with differences between methods tending to be less pronounced. Hence the PB situations labelled 1, 2 and 4 in Figure In order to evaluate the adjustment methods, their performance is assessed through measures of model accuracy (bias), precision (variance) and coverage (type I error rate)  in the s"absolute bias" – the expected difference between the estimated (adjusted) effect and the underlying true effect across all simulations. A negative bias indicates an under-estimate of the true underlying effect, and a positive residual bias indicates an over-estimate of the true underlying effect. Arguably, when interpreting absolute bias, it is important to consider how close the true effect is from the null effect. It is desirable for the adjustment methods to work well near the null where small changes may have impact on the direction of the effect and hence any conclusions. Conversely, if the performance is poorer further away from the null effect (e.g. OR = 3), the consequences are, in some respects, less of a concern because it is unlikely that PB will change the direction of the pooled effect.We consider MSE) which incorporates both the variance and bias of the estimator [In addition, we consider the mean squared error . The underlying true value of θ. The MSE can be thought of as corresponding to the sum of the variance plus the square of bias of Where coverage probability, which can be defined as the proportion of simulations in which the true underlying effect lies within the 95% confidence interval of the predicted effects simulated. This informs how well the type I error is controlled by the statistical model. The coverage probabilities should be approximately equal to the nominal coverage rate to properly control the type I error rate for testing a null hypothesis of no difference in effect size between the true underlying effect and the predicted one; the 5% level is used throughout.We also consider the variance" of the predicted pooled effects. This is used to ascertain the contribution of bias and variance to the MSE.The final measure considered is the average "I2 > 50%. Generally, the methods which allow for heterogeneous data (i.e. include random effects or dispersion parameters) obtain more reasonable coverage probabilities than those which do not.Figures When the underlying effect is increased to OR = 3 figure , greaterτ2 = 0, while varying the number of studies included in the meta-analysis. Figure I2 < 10%) for meta-analyses with typically small numbers of studies (i.e. below 30 studies).Figure Figures The degree of absolute bias is dependent on the underlying odds ratio. This can be explained by p-value induced PB causing 'disfigurement' to the funnel plot which is dependent on the underlying odds ratio; i.e. the funnel plot is almost intact under OR = 3, while a very asymmetrical shape is obtained under OR = 1. The methods that can accommodate heterogeneous data (through the inclusion of random effects or dispersion parameters) are the ones with the most appropriate coverage probabilities among the ones evaluated; these include the Harbord, Peters as well as the Egger-based methods . However, the Harbord and the two Egger (RE-se & D-se) methods report substantially inflated MSE and variance values compared to other methods evaluated.In Figure The conditional regression method, Trim & Fill and standard meta-analysis estimators do not perform particularly well due to low coverage and large amounts of residual bias. As before, for larger values of heterogeneity, the Egger-Var methods  tend to report somewhat lower MSE and coverage probabilities than the Egger ones  thanks to their restrained variance.Also of note is that, under fairly homogeneous effects, adjustment methods based on FE (FE-se & FE-var) and RE (RE-se & RE-var) Egger models provide coverage probabilities well above 95%, implying inappropriately small type I error rates. This can be explained by the inability of these models to accommodate the underdispersion of observed effects (i.e. less variability than would be expected by chance) caused by PB. Conversely, the methods which include a dispersion parameter do not suffer from excessive coverage probabilities due to the fact that they can accommodate underdispersion by allowing the dispersion parameter below the value of one. This means that the Harbord, Peters and the Egger methods (D-se & D-var) perform favourably under scenarios where PB causes underdispersion.I2 exceeds 50% [In this paper we have compared some novel and existing methods for adjusting for publication bias through an extensive simulation study. Results are encouraging, with several of the regression methods displaying good performance profiles. Overall, no particular method consistently outperforms all others. The overall performance of all the methods deteriorates as eeds 50%  and the With respect to the popular Trim & Fill method, we find it hard to recommend over the regression-based alternatives due to its potentially misleading adjustments and poor coverage probabilities, especially when between-study variance is present ,29, althAlthough the standard meta-analysis models are a good approach under lack of PB, they inevitably perform poorly when PB is present. This motivated the examination of regression-based adjustment methods conditional on their associated test for PB. Such an approach is also of interest because it may reflect what is commonly done in practice when dealing with suspected PB. Unfortunately, these conditional approaches did not perform as well as the  alternatives. This may be explained by the fact that all existing tests for PB suffer from low statistical power ,7,25 leaThe persistent low level of coverage probability by the fixed effects Egger models (FE-se & FE-var) under heterogeneous settings render them inappropriate. Equally, coverage probabilities above the 95% threshold produce inaccuracy on the confidence interval; which potentially biases any subsequent assessment of uncertainty around the estimate of interest. This is a serious concern in a decision-making context, where alternative treatments may report similar mean effect sizes. In such cases, accurate quantification of uncertainty to allow discriminating among treatments is vital to facilitate realistic probabilistic statements about, say, cost-effectiveness relative to the alternative treatments. Here, both fixed effects (FE-se & FE-var) and random effects (RE-se & RE-var) Egger models tend to suffer from excessive coverage probabilities under:• Scenarios of underdispersion caused by severe PB figure ;• Mostly homogeneous settings figures , 7, 8, p• Small size meta-analysis figure , provideAdditionally, since in practice it will often be difficult to determine whether heterogeneity is present or not (due to the low power of associated test and distortions caused by PB) this makes appropriate implementation of fixed effect methods difficult.Over the range of simulation scenarios considered, the Harbord, Peters and both Egger dispersion (D-se & D-var) methods would appear to have best overall performance. They do not always produce the least biased estimate, but they do consistently retain good coverage probability levels , while keeping competitive with respect to bias.However, when faced with small size meta-analyses and/or heterogeneity figures , 6, 7, 8One favourable factor in this simulation study is that there was always considerable variation in the sizes of the studies in each dataset. Again, the methods performance will deteriorate if studies sizes are less variable. This is particularly a concern for the regression approaches if all the studies are small, since a larger extrapolation to the intercept would be required.I2 . By doing this, heterogeneity is induced proportionally to the within-study variation. By defining heterogeneity in terms of the I2 statistic means we are focussing on the impact rather than the extent of heterogeneity [τ2) which would lead to an assessment of the methods with respect to absolute degrees of between-study variability. Previous studies that evaluate publication bias methods [In these simulations we defined levels of heterogeneity in terms if ogeneity  across t methods -17,28 haOther methods for PB adjustment are available but were not evaluated in the simulation study. These include a literature on the use of selection modelling techniques . The reaRecently Copas and Malley  presenteSince in medical applications any PB selection mechanisms will be unknown and there will often be too few studies to estimate it from the data, we believe regression-based methods, which make no explicit assumptions about the underlying selection mechanism, may have a useful role in a decision-based context.We believe that a broad range of plausible meta-analyses situations have been evaluated through the scenarios evaluated in the simulation study. And that, given the variability and limited scope of some of the previous simulation studies in the evaluation of methods to address PB in the past, it would be desirable for there to be a consensus simulation framework in which future tests and adjustment methods could be evaluated. To this end, the comprehensive framework developed here could form the starting point for future simulation studies.In conclusion, several regression-based models for PB adjustment performed better than either the Trim & Fill or conditional (regression-based) approaches. Overall, the Egger-var (D-var) and Peters methods are identified as methods with potentially appealing statistical properties for PB adjustment with the Peters method performing better for large odds ratios under the simulation scenarios evaluated. However, it should be acknowledged that while our simulations were extensive, differences in results may be observed if different simulation parameter values were used.Further research is considered worthwhile given our encouraging initial results. To this end, further work exploring the incorporation of information obtained from external sources to form a prior distribution for the regression coefficients is being developed in the hope of improving performance of the regression-based methods.Finally, while we acknowledge that while publication bias is a problem that will not entirely disappear regardless of the statistical method of analysis, ignoring it is an unwise option . We alsoThe authors declare that they have no competing interests. The first author was supported by a Medical Research Council (MRC) Health Services Research Collaboration (HSRC) Studentship. The funding agreement ensured the authors' independence in designing the study, interpreting the data, writing, and publishing the report.AJS and AEA conceived the project and guided SGM in conducting the project. AJS, JLP & SGM designed the simulation study and carried out the statistical analyses. TDS, AEA, KRA & NJC participated in data analysis and interpretation. SGM designed and developed the plots. SGM & AJS drafted the paper, which was later revised by all co-authors through substantial contributions to the contents of the paper. All authors read and approved the final version of the manuscript for publication.The pre-publication history for this paper can be accessed here:http://www.biomedcentral.com/1471-2288/9/2/prepubAdditional results from the simulation study. Additional plots summarising the simulation results for the remaining scenarios.Click here for file


A paradigmatic example of an emotional bias in decision making is the framing effect, where the manner in which a choice is posed – as a potential loss or a potential gain – systematically biases an ensuing decision. Two fMRI studies have shown that the activation in the amygdala is modulated by the framing effect. Here, contrary to an expectation based on these studies, we show that two patients with Urbach-Wiethe (UW) disease, a rare condition associated with congenital, complete bilateral amygdala degeneration, exhibit an intact framing effect. However, choice preference in these patients did show a qualitatively distinct pattern compared to controls evident in an increased propensity to gamble, indicating that loss of amygdala function does exert an overall influence on risk-taking. These findings suggest either that amygdala does contribute to decision making but does not play a causal role in framing, or that UW is not a pure lesion model of amygdala function. People often make suboptimal decisions. A striking example is provided by an influence of irrelevant, emotionally valenced, contexts. In a paradigmatic experimental situation participants are asked to choose between a sure win or loss and a risky gamble. Normatively, such decisions should reflect maximisation of subjective utility, but it turns out these decisions are subject to subtle deviations from optimality that reflect the manner in which choices are framed. Framing a choice as a potential win or a loss biases decision-makers to express a greater proportion of risky choices under the loss frame compared to that under the win frame, a finding termed the ‘framing effect’ . For exaBrain imaging studies implicate the amygdala, orbitofrontal cortex (OFC) and the anterior cingulate cortex (ACC) in the framing effect . ResearcThe ACC is thought to bias decision making towards cognitively efficient strategies by acting as a teaching signal . BecauseThe framing effect has an interesting kinship to Pavlovian-Instrumental Transfer (PIT) effects. PIT refers to the enhancing effect of previously conditioned, but irrelevant, cues on instrumental action vigour and choice . To the Here we set out to test whether the framing effect is impaired in two patients with Urbach-Wiethe (UW), a rare condition associated with bilateral amygdala lesions. Given that this condition has been associated with increased risk-taking behaviours, we have also examined overall gambling frequency in these patients with the prediction that it may be increased relative to controls.11.11.1.1Two German 34-year-old female twins with Lipoid proteinosis (UW) disease who were previously characterized by 1.1.220 age-and-education matched female German controls were recruited from the community. 12 were paid by the hour and 8 were told they would receive the amount they won in the task. We combined them as there were no differences between these control groups, nor interactions with experimental factors. Controls were administered the LPS and the d2 test and performed better on those tests relative to the patients see . The stu1.2sure amount and a gamble. The gamble was presented as a pie chart, showing the chances of winning and losing as portions of the pie, with the win and lose portions coloured green and red, respectively. In half the experimental trials the sure amount was framed as the amount participants would keep of the initial endowment (‘win frame’). In the other half, the sure amount was framed as the amount participants would lose of the initial endowment (‘lose frame’).The procedure replicated a previously reported paradigm  but hereThe expected value of the gamble, namely the magnitude participants could win times the probability of winning, was identical to that of the sure amount. For example, following a €50 endowment a sure win of €10 or a sure loss of €40 may be paired with 20% chance of winning €50, with both options having an expected value of €10. The endowments varied between €25 and €100. For each of the two frames, there were 8 trials with each of four wining probabilities comprising 20%–40%–60%–80%, amounting to 64 experimental trials in total. Thirty-two ‘catch’ trials  were included to verify that participants understood and attended to the task so as to make considered decisions. Half were presented with the win and half with the loss frame. In these trials the sure amount was always half the endowment. In gamble-weighted trials the gamble had markedly higher expected value than the sure amount (95% of winning the endowment), and sure-weighted trials the gamble had a markedly lower expected value (5% chance of winning the endowment). We excluded 6 controls who decided irrationally, preferring the gamble in sure-weighted trials or the sure amount in gamble-weighted trials.2t(19) = 4.13, p = .001. Risk-neutral decisions in this task correspond to 50% gamble frequency. Controls were risk-averse in the loss frame, t(19) = 2.66, p < .05, and did not differ significantly from risk-neutrality in the win frame, t(19) = 1.86, p = 0.08. Z = −.26, BG: Z = −.14). There were no significant differences, t < 1, between the framing effect in each patient and in controls according to the Revised Standardized Difference Test , and under the loss frame in AM . The Crawford t-test (t(20) = 1.45, p = .08; BG: t(20) = 1.66, p = .06; win: AM and BG gambled equally frequently, t(20) = 1.66, p = .06; all tests one-tailed]. This means that in the loss frame an estimated 92% of the healthy population would gamble less than AM and 94% would gamble less than BG; in the win frame, an estimated 94% of the controls would gamble less than the patients.d t-test  showed a2.1t < 1  = 6.78, p < .05 (t-tests showed a non-significant trend in this direction for deciding to take the sure amount in the win frame  and BG was significantly slower from controls when deciding to take the sure amount in the loss frame . All other effects were not significant, p > .1.A 2 (frame) × 2 (gamble choice) ANOVA on reaction time in controls revealed that deciding to gamble was slower than deciding to take the sure amount,  p < .05 ; none of2.3r = 0; d2/framing: r = .19; LPS/gambling: r = .24; d2/gambling: r = −.18, all p > .10).Given the difference between patients and controls in LPS and d2, we examined the correlation in controls between each of these measurements and the framing effect, as well as the overall gambling frequency. None of the correlations were significant (LPS/framing: 3Two UW patients with congenital, complete bilateral amygdala degeneration exhibited an intact framing effect. Patients gambled more frequently when the decision was framed as a potential loss than when it was framed as a potential gain. The influence of the frame was equivalent in patients and controls. This result is surprising because two fMRI studies  using thA positive finding was our observation that across frames, patients gambled more frequently than controls. This can be quantified as an estimate that more than 90% of healthy controls would gamble less than the patients. Our study is therefore consistent with previous results  that patThere are a number of difficulties in directly comparing the aforementioned studies. First, the interaction between domain and group in A critical question is why amygdala lesions should enhance risk propensity. A more likely explanation for patients’ increased risk propensity, we believe, has to do with their ability to learn from feedback, a finding with links to animal models where amygdala lesions are associated with an inflexible coding of associations . In humaThere are two ways to reconcile the present results and imaging data. First, although neuroimaging data show an association between amygdale and the framing effect, this correlation does not imply causation. Disparities between amygdala activation in fMRI and lesion results have previously been raised in the domain of emotion recognition. For example, a number of fMRI studies found that amygdala activation did not distinguish between a range of negative and happy facial expressions . By contSecond, UW is often considered a paradigmatic amygdala disease model. It is nevertheless the case that UW may impact upon the trajectory of development so that UW patients may not be a pure model for loss of amygdala given the possibility of compensation by other brain regions. Animal data support this possibility because neonatal vs. adult-onset bilateral amygdala lesions in monkeys result in different phenotypes. For example, monkeys lesioned at infancy display more fear in social situations , which iIt is a weakness of the study that we did not acquire SCR data, which would have been particularly interesting given the patients’ intact behavioural performance on our task. Because there was no feedback in our task, the differential response to the frame must be due to the value the frame acquired pre-experimentally. It is unclear how amygdala lesions would influence the autonomic response to the frame. Work in animal models shows that amygdala lesions do not abolish all conditioned responses. In humans the pattern has been somewhat mixed, with amygdala lesions attenuating SCR responses to wins and losses  but leavThe influence of amygdala lesions on real-life emotion and decision making has not been explored in detail. A diary study  observedWe conclude that despite anomalous decision making characteristics evident in an increased propensity to risk behaviour in patients with amygdala lesions, these patients nevertheless show an intact framing effect.


Aging is marked by declines in levels of many sex hormones and growth factors, as well as in cognitive function. The P300 event-related potential has been established as a predictor of cognitive decline. We decided to determine if this measure, as well as 2 standard tests of memory and attention, may be correlated with serum levels of sex hormones and growth factors, and if there are any generalizations that could be made based on these parameters and the aging process.In this large clinically based preliminary study several sex-stratified associations between hormone levels and cognition were observed, including (1) for males aged 30 to 49, both IGF-1 and IGFBP-3 significantly associated negatively with prolonged P300 latency; (2) for males aged 30 to 49, the spearman correlation between prolonged P300 latency and low free testosterone was significant; (3) for males aged 60 to 69, there was a significant negative correlation between P300 latency and DHEA levels; (4) for females aged 50 to 59 IGFBP-3 significantly associated negatively with prolonged P300 latency; (5) for females at all age periods, estrogen and progesterone were uncorrelated with P300 latency; and (6) for females aged 40 to 69, there was significant negative correlation between DHEA levels and P300 latency. Moreover there were no statistically significant correlations between any hormone and Wechsler Memory Scale-III (WMS-111). However, in females, there was a significant positive correlation between estrogen levels and the number of Attention Deficit Disorder (ADD) complaints.Given certain caveats including confounding factors involving psychiatric and other chronic diseases as well as medications, the results may still have important value. If these results could be confirmed in a more rigorously controlled investigation, it may have important value in the diagnosis, prevention and treatment of cognitive impairments and decline. A review of the literature reveals that sex hormones have often been associated with changes in behavioral and mental abilities. As noted by Craig and Murphy, estrogen may modulate brain function, and acute loss of ovarian hormones increases neuronal membrane breakdown. Additionally, suppression of ovarian function may reduce activation of brain regions that are critical for memory . Other rThe behavioral effects of the androgens testosterone and dehydroepiandrosterone (DHEA) remain unclear but preliminary reports suggest that their use is associated with improved mood ,5. ThereThe GH/insulin-like growth factor-1 (GH/IGF-1) axis is known to be involved in aging of physiological functions including low levels associating with cognitive decline as a function of age . DeficieSince there is sparse definitive information on hormone plasma levels for measurement of cognitive decline as an age-related process, we decided to investigate this issue.Plasma sex and growth hormones  were analyzed in both males and females across a wide age range and the results were statistically correlated with an electroencephalograph-obtained event-related potential (P300), whose prolonged latency has been established as an accurate predictor of cognitive and memory decline. The methods employed in this study were previously reported by our laboratory . SubjectIn this large  clinically based study, all patients were tested for their response to the P300 event-related potential (ERP) and screened for circulating hormones , and growth hormones IGF-1 and IGFBP-3). Mean ± standard deviation of age for females was 57.4 ± 15.1 and 55.8 ± 14.5 for males.The patients had at least one P300 test. All test interpreters were blinded to other patient results. These patients were selected for study from an outpatient private medical clinical practice  and research foundation in New York City. All subjects signed an approved IRB consent form based on an approval from the PATH Medical IRB committee (registration # IRB00002334). Subjects were made aware that their results could be used in medical research and by signing the informed consent, they volunteered to participate in this study.Venipuncture was done in nonfasting subjects between 8:30 AM and 7:30 PM at baseline examination. Assays were performed blinded to information on the subject. Plasma levels of estradiol and sex hormone-binding globulin were estimated with double anti-body radioimmunoassays . As measures of the levels of bioavailable and free estradiol, testosterone, DHEA, and nonprotein-bound estradiol, respectively, were calculated on the basis of hormone and binding protein levels. For the analysis of growth hormone and insulin-like growth factor binding proteins the laboratory performed standardized procedures. Free testosterone, or unbound testosterone, was calculated by correcting for SHBG and albumin.All patients in this study were analyzed by the Brain Electrical Activity Map (BEAM), a computerized quantitative EEG. A 24-channel EEG recorder was used. Standard techniques were used to measure event-related P300 prolonged latency . The stath percentile) was obtained for each hormone level and used as a cutpoint to define low hormone levels. The hormone data was standardized by assigning a percentile ranking of 1 to 20 to each hormone measure, with 1 representing values ranging from the minimum to the 5th percentile and 20 representing values ranging from the 95th percentile to the maximum. Partial correlations were performed to check for associations between hormone rankings and P300 latency controlling for age. As a consequence of previous analyses, the subject's age plus 300 served as the cutpoint defining prolonged P300 latency for all subjects [As hormone levels are known to be influenced by sex, the analysis was performed for females and males separately. For females, IGF-1, IGFBP-3, estrogen, progesterone, and DHEA were studied. For males, IGF-1, IGFBP-3, free testosterone, and DHEA were studied. Since the study included a broad spectrum of ages, the potential for the age factor to confound the analysis of the complete data set compelled an approach in which subjects were stratified by age decade . This approach helped to negate any influence that age might have on P300 latency and hormone levels, since it has been previously established that the distribution of these measures are not independent of age . Also, lsubjects . Using t negative correlation between DHEA levels and P300 latency. Moreover, for all females, there was a significant  positive correlation between estrogen levels and the number of ADD complaints.While the amplitude of the P300 evoked potential was assessed in this study we did not find any significant associations. However, in terms of latency, significant associations included (1) in males aged 30 to 49, both IGF-1  and IGFBP-3  significantly associated negatively with prolonged P300, and for all males, rank-transformed IGF-1 and IGFBP-3 values were significantly correlated  after partially out the age effect; (2) in males aged 30 to 49, the correlation between prolonged P300 latency and low free testosterone  was significant; (3) in males aged 60 to 69, there was a significant negative correlation  between P300 latency and DHEA levels; (4) in females aged 50 to 59, IGFBP-3 was significantly  negatively associated with prolonged P300 latency; (5) In females aged 40 to 69, there was nearly significant th percentile. We are also cognizant of the relationship of body mass index (BMI) of the patients on hormone levels. In that regard we have already collected data on over 900 patients and evaluated the relationship between growth factors and BMI as well as growth hormone secretion. In future studies we intend to exclude all patients with a BMI of >30. Moreover we are cognizant that many studies only measure hormone levels in the morning to avoid variation as a consequence of diurnal rhythms. This potentially confounding factor must be taken into account as another caveat of the present study. Importantly this fact is particularly true for testosterone.As stated this is a preliminary investigation and future studies will attempt to provide additional required information to make any serious interpretation of the current data. There are many caveats to this study that will provide a blueprint for additional meaningful research. One major problem in this study was the patient group being not well defined due to constraints of the clinical data collection center. While we are cognizant that a perspective study utilizing only healthy volunteers is warranted, the present study does have important value. With that stated our concern is that underlying disease is a major confounding factor because of the presence of a psychiatric disorder (as was the case for the majority of patients in this study). Other clinical disorders of note, such as Alzheimer's disease, diabetes mellitus, and chronic hepatic disease, among others, will affect the P300 measurements. Another caveat is that it is well known that several medications  and also underlying diseases will influence measured hormone levels. Thus in future studies we will attempt to separate patients with chronic diseases and assign categories based on psychiatric disorders, and metabolic diseases where baseline values may be outliers, or further confounded by medication. In addition recent data suggest that head trauma is an important cause of pituitary dysfunction and also a well-known reason for cognitive dysfunction. In future studies the head trauma patients will be excluded from the study group, or at least those that fall into a <25Considering these caveats, our results can be analyzed with caution. In terms of the findings related to the female sex hormone estrogen and the P300 latency results, the literature has not consistently shown that estrogens prevent cognitive decline or Alzheimer's disease. In fact one study showed a small negative effect of higher estradiol levels in memory performance in both females and males and other studies do not support a positive outcome with hormone therapy . In termThe present study also addressed attentional complaints and the potential relation to hormone levels. Follow up to this pilot study on attentional complaints further confirms the early findings. Attention problems have a multimodal dimension. ADHD, including attentional complaints, is related to memory, electrophysiology, and also to genetic and psychiatric factors. Attentional complaints also have Advanced Psychiatric Disease Performance Test (Axis I) sources, such as anxiety disorders, depression, schizophrenia, delusions, etc. It turns out that attention deficits and impaired memory are common to patients with depression . The onlThe present study does not reject the possibility that exogenous estrogens, testosterone, DHEA, and growth hormones may be beneficial in protecting against cognitive decline. To our knowledge this is the first report showing a significant negative association of IGFBP-3, free testosterone and DHEA, with P300 prolonged latency (a predictor of memory impairment and cognitive decline) . In addiThe authors declare that they have no competing interests.ERB developed the original concept and contributed to the development of the data set. TJHC provided input into the clinical protocol and contributed to the development of the manuscript. AC contributed equally to TJHC and provided additional literature references. RLW was involved in editorial assistance as well contributing to the literature. HT was involved in the drafting of the manuscript. making editorial comments. MK was involved in acquisition of funding, collection of data, or general supervision of the research group, JS participated in the design of the study and performed the statistical analysis. KB was involved in the design of the study, drafted the manuscript and was also involved in developing appropriate statistics as well as total overseeing of the project.


A-R are widely prescribed for the treatment of anxiety disorders, epilepsy, and insomnia as well as for pre-anesthetic sedation and muscle relaxation. It has been hypothesized that these various pharmacological effects are mediated by different GABAA-R subtypes. If this hypothesis is correct, then it may be possible to develop compounds targeting particular GABAA-R subtypes as, for example, selective anxiolytics with a diminished side effect profile. The pyrazolo-pyrimidine ocinaplon is anxioselective in both preclinical studies and in patients with generalized anxiety disorder, but does not exhibit the selectivity between α1/α2-containing receptors for an anxioselective that is predicted by studies using transgenic mice.Compounds targeting the benzodiazepine binding site of the GABAin vivo might be influenced by an active biotransformation product with greater selectivity for the α2 subunit relative to α1. One hour after administration of ocinaplon, the plasma concentration of its primary biotransformation product, DOV 315,090, is 38% of the parent compound. The pharmacological properties of DOV 315,090 were assessed using radioligand binding studies and two-electrode voltage clamp electrophysiology. We report that DOV 315,090 possesses modulatory activity at GABAA-Rs, but that its selectivity profile is similar to that of ocinaplon.We hypothesized that the pharmacological properties of ocinaplon A-R modulators.These findings imply that DOV 315,090 could contribute to the action of ocinaplon in vivo, but that the anxioselective properties of ocinaplon cannot be readily explained by a subtype selective effect/action of DOV 315,090. Further inquiry is required to identify the extent to which different subtypes are involved in the anxiolytic and other pharmacological effects of GABA A receptors (GABAA-R) are pentameric membrane proteins that belong to the superfamily of cys-loop ligand-gated ion channels (LGIC), which operate as GABA-gated Cl--selective channels. GABAA-R mediate most of the fast inhibitory neurotransmission in the CNS Ro15–1788  in the presence of diazepam , ocinaplon  or DOV 315,090  for 1 h at 0°C. The samples were then diluted with 5 ml of ice-cold buffer and filtered under vacuum through glass-fiber filters (GF/B Whatman). Filters were washed 3 times with 5 ml of buffer and the radioactivity was quantitated by liquid scintillation counting in 5 ml of Ecolite scintillation fluid (ICN). Non-specific binding determined in the presence of 100 μM Ro 15–1788 (Sigma-Aldrich) was subtracted from total binding to calculate specific binding. Data were analyzed by non-linear regression .For competition binding, 100 μg of membrane protein was incubated in 500 μl of PBS buffer with 0.5 nM  × 100, where I(GABA + Drug) is the current response in the presence of diazepam, and IGABA is the control GABA current. Potentiation data from each oocyte was fitted to the equation Potentiation = Emax × [Drug]/(-pyrimidin-3-yl](pyridin-2-yl)methanone); ocinaplon, (2-pyridinyl pyrimidin-3-yl]methanone); DOV 315,090: (7-(1-Oxidopyridin-1-ium-4-yl)pyrazolo pyrimidin-3-yl)(pyridin-2-yl)methanone, GABA, γ-aminobutyric acid; IGABA: GABA-gated current.cDNA: complementary deoxyribonucleic acid; cRNA: complementary ribonucleic acid; DOV 51892: (7-(2-chloropyridin-4-yl)pyrazolo- [1,5-DB carried out electrophysiological recordings. MCG carried out radioligand binding experiments. EK performed initial electrophysiological experiments. SD developed the data-acquisition hardware and software used in this study. TTG participated in the design of the study, performed the statistical analysis and participated in manuscript preparation. DHF participated in the design of the study and participated in manuscript preparation. PS directed development of ocinaplon at DOV Pharmaceuticals and participated in manuscript preparation. ASB identified major ocinaplon metabolite and participated in manuscript preparation. All authors read and approved the final manuscript.


Normal cellular metabolism is well established as the source of endogenous reactive oxygen species which account for the background levels of oxidative DNA damage detected in normal tissue. Hydrogen peroxide imposes an oxidative stress condition on cells that can result in DNA damage, leading to mutagenesis and cell death. Several potentially significant genetic variants related to oxidative stress have already been identified, and angiotensin I-converting enzyme (ACE) inhibitors have been reported as possible antioxidant agents that can reduce vascular oxidative stress in cardiovascular events.2O2), and the effects were tested by comet assay. Blood samples were submitted to genotyping and comet assay (before and after treatment with H2O2 at 250 μM and 1 mM).We investigate the influences of haptoglobin, manganese superoxide dismutase , catalase (CAT -21A/T), glutathione peroxidase 1 (GPx-1 Pro198Leu), ACE (I/D) and gluthatione S-transferases GSTM1 and GSTT1 gene polymorphisms against DNA damage and oxidative stress. These were induced by exposing leukocytes from peripheral blood of healthy humans (N = 135) to hydrogen peroxide (H2O2 at 250 μM, the GPx-1 polymorphism significantly influenced results of comet assay and a possible association of the Pro/Leu genotype with higher DNA damage was found. The highest or lowest DNA damage also depended on interaction between GPX-1/ACE and Hp/GSTM1T1 polymorphisms when hydrogen peroxide treatment increased oxidative stress.After treatment with HThe GPx-1 polymorphism and the interactions between GPX-1/ACE and Hp/GSTM1T1 can be determining factors for DNA oxidation provoked by hydrogen peroxide, and thus for higher susceptibility to or protection against oxidative stress suffered by healthy individuals. Normal cellular metabolism is well established as the source of endogenous reactive oxygen species (ROS), mainly as a result of normal oxidative metabolism in the mitochondria ,2. BecauUnder normal circumstances, ROS are neutralized by an elaborate antioxidant defense system consisting of enzymes such as catalase (CAT), superoxide dismutase (SOD), glutathione peroxidase (GPx) and numerous non-enzymatic antioxidants ,6],,. In suchSeveral potentially significant genetic variants related to oxidative stress have already been identified ,13. Exam2O2) of leukocytes from peripheral blood of healthy human individuals.Thus, the aim of this study was to verify, through comet assay, the influences of gene polymorphisms on the oxidative damage to DNA. For this, we investigated the genetic polymorphisms of Haptoglobin, MnSOD , CAT (-21A/T), GPx-1 (Pro198Leu), ACE, GSTM1 and GSTT1 against DNA damage and oxidative stress induced by exposure to hydrogen peroxide  and different age groups (17 to 56) randomly recruited in high schools, colleges and universities of Brasília . It was composed of non-hypertensive, non-diabetic, non-dislipidemic and apparently healthy individuals (inclusion/exclusion criterion), of whom 121 were non-smokers and 14 were smokers. Based on the subjects' self-declared skin color, 29.6% were racially mixed, 38.6% were white, 3.7% were black, 2.2% were Asiatic subjects and 25.9% did not declare their color. The volunteers were informed about the purpose of the study and all of them received a random number.This study was conducted according to the guidelines laid down in the Declaration of Helsinki and all procedures involving human subjects were approved by the Ethics Committee for Health Sciences Faculty Research of the University of Brasília, number 092/2008. Written informed consent was obtained from all the registered volunteers and all subjects were free to withdraw at any time during the study.in situ, and blood was then immediately processed for the comet assay.About 5 mL of peripheral blood was collected by venipuncture using Vacutainer tubes with EDTA as anticoagulant for the comet assay and gene polymorphism analyses. Blood collection was done 2O2) at concentrations of 250 μM and 1.0 mM, for five minutes at 4°C. The control slides were prepared under the same conditions, but without the hydrogen peroxide. Slides were then immersed in a freshly prepared cold (4°C) lysis solution  for 1 hour at 4°C. After lysis, slides were placed in a horizontal gel electrophoresis tank with fresh alkaline electrophoresis buffer  and left in the solution for 40 minutes at 4°C. Electrophoresis was conducted at 4°C for 30 minutes at 25 V (0.8 V/cm) and 300 mA. Once completed, slides were washed three times with neutralizing solution , stained with EtBr (20 μg/ml) and analyzed with a Zeiss Axioskop 2 fluorescence microscope  with a total magnification of 400×. One hundred comets on each slide were scored visually by a trained professional as belonging to one of the five classes proposed by Collins et al., 1995 [The comet assay  was performed according to the method developed by Singh et al., 1988 , with a l., 1995 . All expDNA was isolated from the buffy-coat layer by using a Blood genomicPrep Mini Spin Kit  purification kit, and the samples were stored below -20°C until analysis. DNA samples underwent amplification in an MJ PTC-100 . Hp genotypes were determined by allele-specific polymerase chain reaction (PCR) as described by Yano et al., 1998 , while Mclass 1 and 2) and percentage of elevated damages (comets of class 3 and 4) and values of p < 0.05 were considered statistically significant. Besides the influences of gene polymorphisms, the influences of sex (55 men and 80 women), use of tobacco and age groups were also treated statistically. The stratification of subjects according to age was done according to their inclusion in one of the following groups: adolescents (15-19 year-olds), young adults (20-40 year-olds) and middle-aged adults (41-56 year-olds), following the age criteria for reference values of biochemical parameters; for clinical purposes some reference values are different for ages up to 19 years old [The minimum sample size of each genotype was estimated through a power analysis where the maximum tolerable errors were set at 0.10 for the rare genotypes  and at 0.05 for the other genotypes. Statistical analysis was carried out using SPSS  version 15.0. Data were expressed as mean ± SEM (standard error of mean) of the percentage of moderate damages , and the Odds Ratio (OR) with 95% confidence intervals (CI) was calculated to estimate whether genotypes were associated with higher/lower DNA damage. For this, the Pearson chi-square test was applied and differences were considered significant at p < 0.05. To verify differences between the control and treatment slides, the Friedman test followed by the Wilcoxon test were used, respectively for multiple and 2-to-2 comparisons. Differences were considered significant at p < 0.05. The relationship between age groups and DNA damage was analyzed through Pearson correlation test followed by linear regression, and the interactions between two genetic markers in the results of comet assay were analyzed through Multivariate Analyses of Variance.2) test. Values of p > 0.05 indicated HWE. Probability (p) values for co-dominant markers  were generated using Genepopweb Statistical Programme version 4.0 http://genepop.curtin.edu.au. For dominant markers (GSTM1 and GSTT1), χ2 test was calculated using a chi-square calculator program. As the PCR method is not suitable for distinguishing between homozygous  and heterozygous (+/-), these two groups were considered together (non-null genotypes) and compared with the variant group (null genotypes).Allelic and genotypic frequencies were estimated by gene counting, and the goodness of fit of the genotype distribution for Hardy-Weinberg equilibrium (HWE) was tested by chi-square (χ2O2). Significant differences appeared for the comparisons between age groups of 17-19 and 41-56 year-olds (p = 0.021), 20-40 and 41-56 year-olds (p = 0.036), and between the GPx-1 genotypes Pro/Pro and Pro/Leu (p = 0.012) after treatment with H2O2 at 250 μM. Concerning age groups, these significant differences were related to the frequencies of moderate damages between 20-40 and 41-56 year-olds (p = 0.014), and to elevated damages between 17-19 and 41-56 year-olds (p = 0.044) and between 20-40 and 41-56 year-olds (p = 0.008) , Hp1F-1S (N = 4) and Gpx-1 Leu/Leu (N = 5). No significant differences were observed between sexes before (control) and after treatment with hydrogen peroxide HO2. Signi2O2 at 250 μM (p = 0.026). The Odds Ratio (OR) with 95% confidence intervals (CI) was of 0.533 (CI: 0.269 to 1.056) for GPx-1 Pro/Pro (p = 0.070); 2.625 (CI: 1.293 to 5.331) for GPx-1 Pro/Leu (p = 0.007); and 0.468 (CI: 0.151 to 1.450) for GPx-1 Leu/Leu (p = 0.181). No significant differences were observed between the other genetic markers' genotypes or between non-smokers and smokers.There was also a positive significant relationship between age groups and DNA damage after treatment with H2O2 concentrations and DNA-damage levels. The two H2O2 treated groups were significantly more damaged than the controls in almost all analyses , except for Hp1F-1F . The leukocytes treated with 250 μM of H2O2 were also significantly less damaged than those treated with 1.0 mM for the total group , sexes  and age group of 20-40 year-olds . For the genetic markers, the significant differences between treatments with H2O2 at 250 μM and 1 mM were for Hp 1S-2 (p < 0.001), Hp 2-2 (p = 0.044), MnSOD Val/Val (p = 0.012), MnSOD Val/Ala (p < 0.001), CAT AA (p = 0.005), CAT AT (p = 0.022), CAT TT (0.012), GPx-1 Pro/Pro (p < 0.001), GPx-1 Pro/Leu (0.041), ACE DD (p = 0.011), ACE ID (p = 0.002), GSTM1- (p = 0.002), GSTM1+ (p = 0.001), GSTT1+ (p < 0.001) and for GSTM1-T1+ (p = 0.001) and M1+T1+ (p = 0.006).There was a positive correlation between H2O2 (p = 0.027), and for Hp/GSTM1T1 for leukocytes treated with 1 mM of H2O2 (p = 0.041). For the interaction GPx-1/ACE, the greatest damage was presented by individuals carrying GPx-1 Pro/Leu and ACE DD genotypes; while the least damage was observed for individuals GPx-1 Pro/Pro and ACE DD. For the interaction Hp/GSTM1T1, major damages were observed for individuals Hp 1F-1F/GSTM1+T1+ and Hp 1F-1S/M1+T1-, while the smallest damages were seen in subjects carrying Hp 1F-1S/GSTM1-T1- and Hp 1F-2/GSTM1+T1+.When the interaction between two genetic markers was performed, significant results appeared for GPx-1/ACE after treatment with 250 μM of HIS value (-0.4093) compatible with selection in favor of heterozygotes, while for the CAT locus data of Ho, He and FIS value (+0.2141) were compatible with homozygote excess (AA and TT).Results indicated a significant deviation from Hardy-Weinberg equilibrium (HWE) for the CAT p = 0.012), MnSOD (p < 0.001), GSTM1 (p < 0.001) and GSTT1 (p < 0.001) loci  is an important reactive oxygen species (ROS) that arises either during the aerobic respiration process or as a by-product of water radiolysis after exposure to ionizing radiation [2O2 with transition metals imposes an oxidative stress condition on cells. This in turn can result in damage to cell components such as proteins, lipids and principally to DNA, leading to mutagenesis and cell death [2O2 provokes permanent growth arrest and apoptosis in a number of cell types [2O2 at 250 μM, can be more harmful for older subjects. This is in accordance with the literature, since aging is an inevitable biological process and characterized by a general decline in physiological function [The relatively high concentrations of hydrogen peroxide used in the present study were to provide oxidative stress, since this was a secondary aim of this study. The choices of 250 μM and 1 mM were based on a previous result from studies by our group: the treatment with Hg a race . Hydrogeadiation . The reall death . It has ll types . In thisfunction ,37. More2O2 treated groups were not significantly more damaged than the controls. Two previous studies by our group corroborate this suggestion [1 alleles, Hp*1F was the most protective, besides showing that homozygous Hp 1F-1F incurred less DNA damage than other genotypes [locus from HWE was the higher frequencies of Hp1F-1F than the expected [1 alleles can depend on the interaction with GSTM1T1 genotypes.Likewise, since extracellular hemoglobin becomes highly toxic due to the oxidative capacity of its iron-containing heme, which participates in the Fenton reaction to produce reactive oxygen species, and the protective effect of Hp against this oxidative mechanism is phenotype dependent ,21, our ggestion ,38. The expected , indicatexpected , the intAdditionally, synthesis of the antioxidant enzymes is part of the adaptive response triggered by the stress posed by ROS . GlutathIt has been reported that ACE DD polymorphism is a potent risk factor for myocardial infarction . ACE inhIn the same way, the frequencies of variants, such as the Ala allele of MnSOD gene, T allele of CAT gene and Leu allele of GPx-1 gene vary between ethnic groups ,28,45; iThe Brazilian population as a whole is very mixed and heterogeneous, primarily as a result of five centuries of interethnic crosses between Europeans, Africans and Amerindians , and thi2O2 at 250 μM there were significant differences between DNA damage in younger and older age groups, indicating that oxidative stress can be more harmful for older subjects. Under similar conditions, GPx-1 polymorphism significantly influenced results of the comet assay, and a possible association between Pro/Leu genotype with higher DNA damage was found. Under increased oxidative stress due to hydrogen peroxide treatment, the highest or lowest DNA damage also depended on the interaction between GPX-1/ACE and Hp/GSTM1T1 polymorphisms. This indicates that these polymorphisms and interactions can be determining factors for DNA oxidation provoked by hydrogen peroxide, and thus for higher susceptibility to or protection against oxidative stress suffered by healthy individuals.After treatment with H2O2: hydrogen peroxide; ROS: reactive oxygen species; CAT: catalase; MnSOD: manganese superoxide dismutase; GPx-1: glutathione peroxidase 1; Hp: haptoglobin; ACE: angiotensin I-converting enzyme; GST: glutathione transferase; SEM: standard error of mean.HThe authors declare that they have no competing interests.All authors have made a substantial contribution to this study and to the writing and editing of the manuscript. Additional contributions are as follows: ALMV, CKG and MNKG were responsible for the conception and design of the study; CKG and MNKG coordinated this study; ALMV, PCZA, AKA and GSL were responsible for the genotyping and comet assay; CAG was responsible for the statistical analysis of data and ALMV for the interpretation of data as well as for structuring and writing the manuscript. All authors read and approved the final manuscript.