Diabetes can affect many parts of the body and is associated with serious complications. Oxidative stress is a major contributor in the pathogenesis of diabetic complications and bilirubin has been shown to have antioxidant effects. The number of studies on the effect of bilirubin on the risk of diabetic complications has increased, but the results are inconsistent. Thus, we performed a meta-analysis to determine the relationship between bilirubin concentration and the risk of diabetic complications, and to investigate if there was a dose-response relationship. We carried out an extensive search in multiple databases. A fixed or random-effects model was used to calculate the pooled estimates. We conducted a dose-response meta-analysis to analyze the association between these estimates. A total of 132,240 subjects from 27 included studies were analyzed in our meta-analysis. A negative nonlinear association between bilirubin concentration and the risk of diabetic complications was identified , with a nonlinear association. We also found that there was a negative association between bilirubin concentration and the risk of diabetic nephropathy, diabetic retinopathy and diabetic neuropathy. The results of our meta-analysis indicate that bilirubin may play a protective role in the occurrence of diabetic complications. Diabetes Mellitus (DM) is an important non-communicable disease, and is a serious threat to human health and global economies. Due to population ageing, urbanization and lifestyle changes, the number of people with DM has increased sharply in both developed and developing countries1st century.DM has serious effects on many parts of the body, such as the eyes, kidneys, limbs and nervous system. Therefore, diabetic patients can develop many microvascular complications, such as blindness, kidney failure, and lower-limb disorders leading to amputation. From 2005 to 2008, 4.2 million diabetic patients aged 40 years or older developed diabetic retinopathy (DR), which is caused by damage to the small blood vessels in the retina2+) due to the effects of heme oxygenase (HO), which is an enzyme system comprised of two forms: HO-1 and HO-2. Biliverdin is subsequently reduced to bilirubin by biliverdin reductase. Bilirubin concentration increased as higher HO-1 expression increases1112et al. demonstrated that the prevalence of vascular complications was lower in diabetic patients with Gilbert’s syndrome than that in diabetic patients with normal bilirubin concentrationAs one of the most highly conserved groups of non-polar molecules, bilirubin has been shown to have antioxidant and anti-inflammatory effects7Long-term hyperglycemia leads to the generation of reactive oxygen species, mainly by mitochondria. The reaction process occurs in endothelial cells, and led to vascular complications. In diabetic patients, bilirubin may exhibit its potent antioxidant property by inhibiting lipid peroxidation and attenuating low-density lipoprotein (LDL) oxidation18192022Potential studies 2323) were identified from three electronic databases  using search strategy. After reading the title and abstract, we found that 412 studies were duplicates and 1413 studies did not report the relationship between bilirubin concentration and the risk of diabetic complications. Four hundred and ninety-eight studies were read in full, and 471 studies were excluded for various reasons. In total, 27 studies7181920212223 were i2425262728293031323334353637383940414218192426272930333435402023253136212237383971923312021222530323334353738394041421824262728192223242529313233401819252627283034353639212. The proportion of smokers in the study population ranged from 2.60% to 48.07%, and the proportion of alcohol drinkers in the study population ranged from 7.80% to 51.85%. The mean of duration of DM ranged from 6.14 to 15.19 years.We assessed 13 biochemical indices which may be related to the occurrence of diabetic complications. We assessed the quality of included studies using the quality assessment toolet al.et al.et al.et al.et al.et al.et al.2 = 87.7%, P < 0.001)  . In ordeIn the dose-response analysis, we found that 23 sets of statistical data from 13 of the included studies were suitable. A nonlinear association between bilirubin concentration and the risk of diabetic complications was found (P < 0.001). A degree of heterogeneity was observed (P = 0.032). In the sensitivity analysis, we omitted one study at a time, and the pooled OR ranged from 0.76 (95% CI: 0.72–0.80) to 0.79 (95% CI: 0.76–0.83). The results of the sensitivity analysis showed that the results for the effects of bilirubin on diabetic complications were robust.Both the Begg’s and Egger’s tests showed that publication bias existed in the comparison between bilirubin concentration and the risk of diabetic complications . Thus, we performed the trim and fill method to identify and correct the asymmetry of the funnel plot, and the results showed a negative association between bilirubin concentration and the risk of diabetic complications .2 = 88.1%, P < 0.001). We also used subgroup analysis with dichotomous variables and the results show that there was a significant association between bilirubin concentration and the risk of DN. The overall and subgroup results are shown in We extracted 13 statistical data sets from 11 studies on DN. We also found that there was a negative association between bilirubin concentration and the risk of DN , with high heterogeneity . We also used subgroup analysis with dichotomous variables and the results show that there was a significant association between bilirubin concentration and the risk of DR. The overall and subgroup results are shown in We extracted 13 statistical data sets from 12 studies on DR. We found that there was a negative association between bilirubin concentration and the risk of DR , with moderate heterogeneity . As the number of studies on diabetic neuropathy was small, subgroup and sensitivity analyses were not performed.In this study, diabetic neuropathy included DPN and DCAN. We found that there was a negative association between bilirubin concentration and the risk of diabetic neuropathy , with moderate heterogeneity , and similar results were also found for DN , DR  and diabetic neuropathy . The dose–response relationship between bilirubin concentration and the risk of diabetic complications was non-liner. Bilirubin showed a protective effect on the risk of diabetic complications.vivo study on subjects with Gilbert’s syndrome had found that there was a negative association between serum bilirubin concentration and soluble forms of CD40 ligand and P-selectin13474950Bilirubin has potent antioxidant properties, which were first reported in 1954et al. showed that serum bilirubin concentration was negatively correlated with the level of oxidative stress and positively correlated with the levels of antioxidative enzyme such as superoxide dismutase, catalase, and glutathione peroxidase193843Diabetes can affect many parts of the body and is associated with serious complications, such as heart disease, stroke, blindness, kidney failure, and lower-limb disorders leading to amputation. Oxidative stress is a major contributor in the pathogenesis of diabetic complicationset al.et al.et al.For example, in DN, Hamamoto In order to ensure that the results of our meta-analysis were both reliable and credible, we carried out the following ways: First, we extracted 38 results on the association between bilirubin concentration and the risk of diabetic complications from the included 27 studies, which helped us to understand the effects of bilirubin on the various complications of DM. The ORs which were extracted from the included studies, were adjusted to take into account the common confounders , this ensured the stability of the results of our meta-analysis. Second, the standardized NOS was not suitable for assessing the effects of bilirubin on diabetic complications, therefore, we modified the NOS to satisfy our assessment as the included studies were evaluated more accurately, and this ensured the credibility of the results of our meta-analysis. The included studies did not have a low quality assessment rating, and the results of subgroup analysis were similar to the overall results. Third, we not only analyzed the association between bilirubin concentration and the risk of diabetic complications, but also determined the dose-response relationship, and found that there was a nonlinear association between bilirubin concentration and the risk of diabetic complications. We further analyzed the data for DN and DR. The results were similar to the overall results. This ensured the reliability of the results of our meta-analysis. Fourth, the potential confounders that were extracted contained dichotomous and continuous variables, thus we respectively carried out subgroup analysis and meta-regression, to analyze the effects of the potential confounders on the results of our meta-analysis. We found that FBG and HbA1C may be the source of heterogeneity. Blood glucose control reduced the risk of diabetic complications, especially microvascular  disease2 in the population-based group was lower than 50%. Third, publication bias was observed, and in order to eliminate this, the trim and fill method was used, and a negative association between bilirubin concentration and the risk of diabetic complications was observed. Fourth, we found that only one study had been carried out in Australia, and the other studies were carried out in Asia. This might lead to selection bias due to race. There was a significant difference in the association between bilirubin concentration and the risk of DA in the Australian study, these results suggested that bilirubin may have an impact on the risk of diabetic complications not only in Asian populations, but also in populations from other counties. We also found that there were only five cohort studies, which accounted for merely 18% of the included studies. Thus, large-sample, long-term cohort studies are needed to validate our results, especially studies from countries outside Asia.Although we attempt to ensure that our results were reliable and credible, there were some limitations in our study. First, because the authors of this article did not have access to EMBASE, we could not directly search the relevant articles in this database. To make up for this limitation as much as possible, we searched the relevant articles in Google Scholar. If we found the relevant article in Google Scholar, we purchased the article or sought help online. Second, the source of the subjects in 80% of included studies was hospital-based, these studies may have a certain degree of selection bias as no randomization was performed. We performed subgroup analysis on the source of subjects, and found a negative association between bilirubin concentration and the risk of diabetic complications, and IIn summary, we collected literature that studied the association between bilirubin concentration and the risk of diabetic complications. Our meta-analysis indicated that higher bilirubin concentration reduced the risk of diabetic complications. This meta-analysis was the first comprehensive quantitative assessment of bilirubin on diabetic complications, suggested that bilirubin may be used as a biomarker of the occurrence of diabetic complications.An extensive search strategy was carried out in multiple databases by October 27, 2016 . In order to collect the relevant articles as many as possible, we set the following key words: bilirubin and (“retinopathy” or “nephropathy” or “neuropathy” or “vascular complication”). We also reviewed reference lists from all the relevant original research and review articles to identify additional possibly eligible studies. There is no language restriction.According to the inclusion criteria, the possibly relevant articles which had been collected were independently assessed by three reviewers . Any disagreements were solved by discussion.The diagnostic criterion for DM, prediabetes and hyperglycemia was used by the World Health Organization (WHO) 1999 criteriaThe included studies in our meta-analysis met the following criteria: (1) the study should investigate the relationship between bilirubin concentration and the risk of diabetic complications; (2) the study should contain the odds ratio (OR) with 95% confidence intervals (CIs) on the relationship between bilirubin concentration and the risk of diabetic complications, or had provided available data to calculate the corresponding estimate; (3) if the results of more than two study came from the same population, the latest or highest-quality result was adopted.The studies were excluded from our meta-analysis if the study did not investigate the relationship between bilirubin concentration and the risk of diabetic complications. The exclusion criteria also contained: (1) the study was not original research, for example editorial, commentary and review; (2) subjects in the study did not involve humans, for example animal experiments, chemistry and cell-line studies; (3) individual case reports; (4) the study did not provide sufficient data to calculate the corresponding estimate on the relationship between bilirubin concentration and the risk of diabetic complications.The included studies were independently read in detail by two reviewers (Bo Zhu and Xiaomei Wu). Two reviewers extracted the information by a table, which was prepared using Microsoft Excel 2007. The information extracted included the following: author, year of publication, country; study design, source of subjects, number of subjects and baseline risk factors (body mass index (BMI), Duration of diabetes, Smoking and alcohol consumption), outcome and its definition, biochemical indexes (fasting blood glucose (FBG), hemoglobin A1c (HbA1C), hypertension, systolic blood pressure (SBP), diastolic blood pressure (DBP), dyslipidaemia, total cholesterol (TC), triglycerides (TG), high density lipoprotein-cholesterol (HDL-C), low density lipoprotein-cholesterol (LDL-C) and serum uric acid (SUA)), the results of statistical analysis. In order to reduce the effects of confounding factors on the pooled estimates, we preferred to extract and analyze the adjusted estimates rather than unadjusted estimates. When the studies only provided categories of bilirubin concentration, the results of the highest category were selected for analysis. For studies, which presented several estimates adjusted for different number of potential confounders, the estimate that adjusted for the most number of potential confounders was selected for analysis.The units of bilirubin concentration were often used as μmol/L and mg/dL. In order to make unit consistency in all included studies, we converted mg/dL to μmol/L with multiplying by 17.1.The quality of final included studies was independently assessed by two reviewers (Bo Zhu and Xiaomei Wu). There was no standardized quality assessment tool on assessing the effects of bilirubin on diabetic complications, so NOS was modified to satisfy our requirement according to previous study2, which ranged from 0% to 100% . When I2 was larger than 50%, a DerSimonian and Laird random-effects model was used; otherwise, the fixed-effects model was used. If there was highly heterogeneity between studies, we used subgroup analysis and meta-regression to find the source of heterogeneity. We performed subgroup analysis by country, number of subjects, design, source of subjects, FBG, HbA1C, the study quality. Restricted maximum likelihood (REML) based random effects meta-regression was carried out to investigate the source of heterogeneity among studies according to the relevant factors. We also used the sensitivity analysis to evaluate the robustness of the results in our meta-analysis. In the sensitivity analysis, we excluded each study in turn and obtained the pooled estimates from the remaining studies. The purpose of sensitivity analysis was to evaluate the effect of a single study on the overall pooled estimates.In our meta-analysis, we used the pooled odds ratios (ORs) and 95%CIs to analyze the relationship between bilirubin concentration and diabetic complications, which were calculated by using the inverse variance-weighting method. The Chi-square-based Q-test was used to evaluate the heterogeneity among the individual studies. Heterogeneity was quantified based on Iet al.et al.We conducted the dose–response meta-analysis to calculate study-specific slopes  and 95%CIs, which proposed by Greenland S. The possibility of publication bias was assessed using the Begg and Egger’s test. We also performed the Duval and Tweedie nonparametric “trim and fill” procedure to further assess the possible effect of publication bias in our meta-analysis. A two-sided P value < 0.05 in statistical process was considered significantly different. All statistical analysis was performed with the Stata software package .How to cite this article: Zhu, B. et al. Effect of bilirubin concentration on the risk of diabetic complications: A meta-analysis of epidemiologic studies. Sci. Rep.7, 41681; doi: 10.1038/srep41681 (2017).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


Poisoning, a subtype of physical injury, is an important hazard in children and youth. Individuals with ADHD may be at higher risk of poisoning. Here, we conducted a systematic review and meta-analysis to quantify this risk. Furthermore, since physical injuries, likely share causal mechanisms with those of poisoning, we compared the relative risk of poisoning and injuries pooling studies reporting both. As per our pre-registered protocol (PROSPERO ID CRD42017079911), we searched 114 databases through November 2017. From a pool of 826 potentially relevant references, screened independently by two researchers, nine studies  were retained. We pooled hazard and odds ratios using Robust Variance Estimation, a meta-analytic method aimed to deal with non-independence of outcomes. We found that ADHD is associated with a significantly higher risk of poisoning . Results also indicated that the relative risk of poisoning is significantly higher than that of physical injuries when comparing individuals with and without ADHD . These findings should inform clinical guidelines and public health programs aimed to reduce physical risks in children/adolescents with ADHD. Poisons can be inhaled, ingested, injected or absorbed. On a global scale, poisoning is estimated to cause 350,000 deaths every year, of which 45,000 refer to individuals under the age of twenty1, and, more generally, it leads to higher mortality and morbidity rates in this age group6.Poisoning is defined by the World Health Organization as “an injury that results from being exposed to an exogenous substance that causes cellular injury or death”5. The relationship between age and risk of poisoning has a bimodal distribution with two peaks of highest risk between the ages of 1 and 4, as well as between 13 and 18 years of age5. Age is also associated with a change in the mechanism of poisoning: whereas most poisonings before the age of fourteen are unintentional, the proportion of intentional poisonings increases dramatically from that age onward7. The majority of suicide intents in adolescents consist of intentional intoxications6, accounting for one third of total poisonings in that age range5.Risk factors for poisonings include age and sex, among others. Being male is related to a higher poisoning risk across all age groups8, which makes it the most frequent neurodevelopmental disorder in children and adolescents. It is characterized by inattentive and/or hyperactive- impulsive symptoms that have a negative impact on social9, academic10, and health domains13, and reduce the quality of life14. ADHD is approximately four times more common in boys than in girls. Pharmacological treatment, including psychostimulants (methylphenidate and amphetamines) and non-psychostimulants , is an important component of the multimodal treatment of ADHD15. In addition, a high percentage of patients have comorbid disorders hence increasing the probability of patients being poly-medicated16.Attention-deficit/hyperactivity disorder (ADHD) has a world-wide estimated prevalence of around 5%17. Therefore, a plausible hypothesis is that ADHD symptoms  could lead to a similar increase in the risk of poisoning. Impulsivity might be an important factor, especially considering that it is significantly associated with suicide attempts, as shown in a recent meta-analysis18. Indeed, a recent systematic review on the relationship between ADHD and suicide concluded that there is a positive association between ADHD and suicidality in both sexes and in all age groups that was likely mediated by the presence of comorbid disorders19. Furthermore, individuals with ADHD frequently have more access to potentially harmful medications that many of them take either for the disorder or for its comorbidities.A recent meta-analysis by our group has demonstrated that the risk of physical injuries is significantly higher in children and adolescents with ADHD compared to the typically developing population. Additionally, this risk is significantly reduced by the use of ADHD medications21. Therefore, a meta-analysis quantifying the risk of poisoning in children/adolescents with ADHD is timely. Of note, the previous meta-analysis on the risk of physical injuries excluded studies that specifically analyzed the risk of poisoning17. To fill this gap and complement the previous meta-analysis, we conducted the present meta-analysis aimed to quantifying the pooled risk of poisoning in children/adolescents with ADHD compared to non-ADHD controls. A secondary aim was to compare the magnitude of the risk of unintentional physical injuries and poisoning from studies that reported both. We hypothesized that children and adolescents meeting criteria of ADHD would have significantly higher rates of poisoning compared to those without ADHD, and this increased probability of poisoning would be greater than that of physical injuries.Whereas a higher rate of poisoning in children and adolescents with ADHD in comparison with their typically developing peers has been reported in individual studies, the magnitude of the association is unclear29. Articles that were considered possible candidates for inclusion during the first screening stage but were later deemed ineligible when the full text was assessed are listed in Table 32, similarly to the Clinical Practice Research Datalink and Hospital Episode Statistics (CPRD-HES) from the United Kingdom33. The full process of article search and selection is shown in the PRISMA flow diagram in Fig. Searches carried out in 114 databases  in November 2017 led to 826 articles, whereas nine studies were included in the final stage of the systematic review and meta-analysis21 were based on large epidemiological databases. More specifically, there were two regional28 and three national databases26, a nationally representative survey27, a population-based prospective cohort24, a study using administrative claims from a self-insured company29, and a case-control study using a hospital-based registry21. Therefore, all studies but two27 analyzed administrative databases not specifically designed for research purposes at their inception34. A strength of the included studies is that they tended to have large sample sizes . The systematic review and meta-analysis pooled data from a total sample of 84,756 and 1,398,946 children and adolescents with and without ADHD respectively.The origin of the studies was varied, comprising North-America, Europe, Asia and Australia. All studies but oneOverall, poisoning cases were uncommon. The median number per study of poisoned individuals that suffered from ADHD was 14 (range 2–184), whereas the median number per study of poisoned individuals who did not suffer from ADHD was 29 . Prevalence (per 1000) ranged between 3.5 and 60 (median 16) in children and adolescents with ADHD and between 0.8 and 37.3 (median 4.8) in children and adolescents without ADHD.28. This is a probable cause for the much higher risk of poisoning in both the group with and without ADHD in this study. The retrospective nature of studies and the use of administrative databases were also related to the type of strategies used to identify cases with ADHD and to define poisoning. In the majority of studies, ADHD diagnosis was defined based on ICD codes at visit discharges29, by taking medications for ADHD22, or based on the combinations of the two. However, two studies27 used scales of symptoms and DSM criteria. Similarly, diagnoses of poisoning were defined based on ICD codes. Whereas most studies included poisoning from medicinal and non-medicinal origins, two studies25 only included poisoning from medicinal drugs. Specifically, one25 analyzed poisoning cases that were intentional in nature, which led to a much smaller prevalence of poisoning in in adhd and controls.The ranges of ages of poisoning were large in most cases. An exception was a study in which ADHD was diagnosed in school-age children but retrospectively considered the risk of poisoning during pre-school2 index35, was high . Risk of small sample bias was not significant according to Begg’s adjusted rank correlation and Egger’s test  of poisoning between adolescents with and without ADHD, included eleven outcomes derived from nine studies. Variation among effect sizes was important as they ranged between 1.2 and 7.98 median 3.47). The overlap between confidence intervals (CIs) was small. The lower bound of the CI ranged between 0.5 and 3.58 (median 1.91) and the upper bound between 2.6 and 49.35 (median 5.64). All analyses were carried out using Robust Variance Estimation (RVE) to take into account dependence between outcomes. Individuals with ADHD had a significantly increased risk of poisoning compared to individuals without the disorder  as shown in the forest plot in Fig. .47. The 2 = 0%). The average RR was 2.59  when only odds ratios were taken into account. Similarly, the pooled RR changed to 3.01  when only statistically adjusted RRs were entered in the meta-analysis, and increased to 5.62  when uncontrolled effect sizes were used. Since only two studies did not combine poisoning and intoxication cases (defined using the ICD codes), we could not carry out a sensitivity analysis including only studies that focused on a strict definition of poisoning. It must be noted that in the case of crude effect sizes and hazard ratios only three and four studies were respectively included in the analyses and, therefore, confidence intervals with RVE are unreliable36. Changing the p parameter within RVE, a value that accounts for the correlation between outcomes within studies, did not change the previously stated estimation of effects.In general, results were relatively robust to sensitivity analyses. When only studies using hazard ratios as effect measure were included, the resulting average RR was 3.91 .Regarding the risk of bias, ratings on the Newcastle-Ottawa Scale (NOS) tended to be high . The items of the scale used in the present meta-analysis, subscores and total score for each study can be found in Table Similarly, the sub-group analyses carried out in order to assess the effect of age were not statistically significant: results of the between-studies comparison of outcomes in children under ten years old against outcomes in which age was not specified were B = 0.299, 95% CI = −0.404 to 1.000 p = 0.279; when outcomes from participants with unspecified age were compared to outcomes obtained from participants over 10 years, results were not significant ; finally, when outcomes from individuals under and above 10 were compared, results were not significant either .21 reported only poisoning outcomes. The outcomes related to physical injuries  are summarized in Table 2 = 64.72, df = 7, p < 0.001, I2 = 89.2%). Risk of small sample bias was not significant . The relative risk of physical injuries was significantly smaller than the one in the case of poisoning . A forest plot comparing the two combinations of outcomes, with effect measures from studies which reported both physical injuries and poisoning outcomes, is shown in Fig. An important question of our systematic review and meta-analysis was whether the relative risk of poisoning was statistically different from the relative risk of suffering physical injuries in general. In order to answer this question, we extracted effect sizes reporting the relative risk of unintentional injuries from studies that reported both. Eight studies reported outcomes for both poisoning and physical injuries and one study17. The same occurred when the analysis was limited to studies reporting hazard ratios, as the confidence intervals reported in our previous article for the pooled estimates of physical injuries  did not overlap with the confidence intervals for the overall effect of poisoning 17. This is further evidence to support that the relative risk of poisoning is significantly greater than the relative risk of suffering physical injuries in children and adolescents with ADHD.The confidence intervals of the pooled estimates of physical injuries reported in the present meta-analysis , and those in our previous article , did not overlap with the confidence intervals for the overall effect of poisoning Poisoning is an important cause of morbidity among children and adolescents worldwide. Children with ADHD may represent a particularly vulnerable group, but so far, no pooled estimates of this risk were available. The present meta-analysis has concluded that there is a significantly higher risk of poisoning in children and adolescents with ADHD compared with their non-ADHD peers, with an estimated relative risk of 3.14 (95% CI = 2.23 to 4.42). Taking into account the prevalence of ADHD, the disorder could be a major factor contributing to the overall number of pediatric poisonings. Results derive from the combination of nine large population-based studies with a combined sample close to 1.4 million children and adolescents from the general population and 85k individuals with ADHD. The confidence interval of the estimate is quite narrow (95% CI = 2.23 to 4.42) because of the large sample size.38. Specific features of the disorder such as impulsivity and inattention are likely a major risk factor for poisoning. Our work provides solid meta-analytic evidence further highlighting that ADHD is a disorder with consequences that are not limited to the behavioral or educational domains. It has an impact on different health aspects39 and hence, leads to a reduction in the overall quality of life of the patients and their families14.This increased risk is consistent with previous research, showing a significantly higher risk of physical injuries in individuals with ADHDResults were generally robust to different sensitivity analyses. Furthermore, there was no statistical indication of small-sample bias, including publication bias. However, these analyses were limited by the final number of outcomes and studies in our meta-analysis, so that our results could partly reflect a lack of statistical power. A possible qualitative interpretation of the funnel plot related to our analyses is that studies with greater standard errors  were more likely published if they showed an increased risk of poisoning in children/adolescents with ADHD. If real, hidden reporting bias could be leading to an overestimation of the overall effect. However, a line of evidence makes this overestimation unlikely. In most published studies, relative risk of poisoning was only a secondary outcome among several other subgroups of injuries, whereas the central finding of these studies was an overall increased risk of injuries. In such case, publication bias due to results of only one of many secondary outcomes is unlikely. Moreover, within study comparisons are robust to publication biases. If there was an overall publication bias towards a higher risk of injuries in ADHD (and not only of poisoning), our within-study analytic strategy demonstrating that the relative risk of poisoning is greater than the relative risk of physical injuries in general would not be affected.34 are also limited in number and in many cases will not include measures of interest for the specific research question. Both issues, number of studies and measures reported in them, limited our analyses. We were not able to carry out possibly informative regression analyses, such as evaluating a relationship between the risk of poisoning and ADHD symptoms, or controlling for comorbidity. Similarly, the statistical power of our regression analyses was also very low. Indeed, as it is always the case in frequentist statistics, a lack of significance only indicates that there is not enough evidence for an effect, which does not necessary involve that there is not such an effect. Given that our statistical power was limited due to the characteristics of existing research, our results leave an open door for future studies evaluating such effects.As already mentioned, whereas the pooled sample in our meta-analysis was very big, the number of studies was limited. This can be explained by the fact that poisoning cases are a rare event and, therefore, very large databases are typically needed to be able to carry out epidemiological studies on this outcome. These large epidemiological studies are difficult to conduct: on the one hand, prospective cohorts are expensive and limited; on the other hand, administrative databases 2 statistic (64.4% when only OR were included and 0% when only HR were included). However, this difference between results cannot be explained by the different outcome measure per se but from other study characteristics that co-occur with the selection of the outcome measure. Taking into account the fact that studies reporting HR tended to be larger, more representative of the population, had better statistical control of possible confounders, and the heterogeneity of their results was smaller, the estimation obtained when only including HR studies might better reflect the true relative risk of poisoning in children and adolescents with ADHD. It must be noted, however, that only four studies were included in the case of HRs and hence the confidence intervals calculated through RVE could be wider than expected36.It is also noteworthy that despite our strict inclusion criteria, heterogeneity among studies was significant. Heterogeneity was dealt through the use of random-effects models that assume that the true effect size might differ from study to study, and that studies included in the analysis are a random sample of all possible studies that meet the inclusion criteria. Additionally, we investigated the origins of heterogeneity through meta-regression. The most influential factor driving heterogeneity was the type of outcome measure (Hazard or odds ratios), which influenced not only the pooled effect size  but also the I40, and they increase with an easier access to pharmacological drugs. Indeed, access to medications has been reported as a risk factor for unintentional poisoning42. Children and adolescents with ADHD have more access to medications than developmentally normal individuals do. Nearly 60% of ADHD diagnosed children receive pharmacological treatment with stimulants and other drugs44, and in many cases a single individual will be prescribed several formulations of the same medication45. In addition, many individuals will receive additional medications for comorbid disorders16, such as oppositional defiant disorder, conduct disorder, anxiety, coordination problems, depression, tic disorders and Tourette syndrome. Hence, the poly-pharmacy status in many ADHD patients could increase the likelihood of an accidental poisoning. Moreover, comorbid mental disorders might make children and adolescents with ADHD even more prone to an accidental poisoning. Data on comorbidity was not reported in most of the studies included in the present meta-analysis, so that the impact of comorbidity could not be elucidated. Further research should clarify whether medicated individuals are at a greater risk of poisoning, if comorbidities increase the risk of intoxications, and to which extent these effects can be disentangled. The role of ADHD medication is even more complex. ADHD pharmacological treatment has already been shown to reduce the risk of suffering an unintentional injury17, and conversely, it reduces the risk of driving accidents in adults47. Drugs used to treat ADHD could have a similar effect on the risk of poising: as they improve attention and impulsivity, they could lead to a reduction in the risk of poisoning. However, our systematic review was not able to find any studies on the effect of medication on poisoning risk. Therefore, the relationship between medication effects and risk of poisoning in ADHD deserves further clarification.An additional major finding of the meta-analysis is that the relative risk of poisoning in individuals with ADHD compared to individuals without it was statistically higher than the overall relative risk of physical injuries. Eight studies reported both injuries in general and poisoning cases, hence permitting a within-study evaluation of the effect of type of injury. Of note, the risk of injuries from the combination of the eight studies was 1.54 (95% CI = 1.33 to1.78), closely matching the results from our previous meta-analysis on the risk of physical injuries, in which the mean RR was 1.53 (95% CI = 1.40 to 1.67). Several factors could be accounting for this increased relative risk. Accidental overdoses due to a difference between the taken and prescribed doses are common in pediatric populations5. As the child grows, there is an increase of intentional poisonings, although the total percentage of intentional poisonings remains lower than the percentage of unintentional cases5. For the second age group, recreational drug usage and suicide attempts are important causes of poisoning. Regarding the specific case of individuals with ADHD, adolescents with the disorder use more frequently drugs recreationally48, including their own medications49. For example, a study carried out among adolescents and young adults with the disorder reported that 14.3% of the participants in the study had once abused of their prescribed pharmacological treatment50. Furthermore, evidence tends to support the fact that individuals with ADHD have a higher risk of suicide and suicide attempts19. In summary, ADHD adolescents could be especially prone to intentional (suicide attempts) or semi-intentional  cases of poisoning, compared to younger children with the disorder and also the general population, and this could be driving in part the higher risk of poisonings compared to physical injuries. However, a direct test of this hypothesis was not possible in our meta-analysis since included studies did not differentiate between intentional and unintentional poisonings. Whereas future studies should try to address this issue, it must be noted that this differentiation is likely impossible when using administrative databases. We sought for indirect support for the role of intentional poisoning through a meta-regression including age as a covariate, but the results of this meta-regression analysis were not significant. Therefore, the role of age as a mediator in this issue still remains unknown.In terms of age effects, poisoning incidence has two peaks across the child life span. The first peak occurs in the first years of life and the second is around the beginning of adolescence, changing also the causal factors of poisoningThe results of our systematic review/meta-analysis should be considered in the light of its strengths and limitations. As for the strengths, we pre-registered the protocol in a publicly available repository (PROSPERO), reducing the risk of reporting bias. Furthermore, we endeavored to perform a comprehensive and systematic search in several databases, with no restrictions in terms of language, date, or document type. Additionally, we used a state-of-the-art tool, the Newcastle-Ottawa scale, to assess the quality of the retained studies. Furthermore, the included studies typically used big longitudinal or administrative cohorts or national surveys, which provide adequate statistical power to estimate the overall incidence of an infrequent type of event, such as is poisoning. There are also a number of limitations that should be taken into account, which are mostly related to the individual studies that we included rather than to methodological issues with our systematic review/meta-analysis. First, intentionality of poisoning was not controlled in the included studies. Second, we could not find sufficient data to evaluate the effects of age, medication status or comorbidities on the risk of poisoning in ADHD. Since these major confounders were not controlled for in our analyses, the increased relative risk cannot be directly ascribed to ADHD. Although our results support the conclusion that individuals with ADHD in the real world suffer more poisoning events than those without it, we cannot know what factor or factors are at the origin of this relationship and in this regard, any causative explanation derived from them should be taken with caution.1, whose treatment involves a huge cost of economic and human resources52. The present meta-analysis has shown that children and adolescents with ADHD are a population with an increased risk of poisoning. Specific preventive measures in this population could help to minimize this risk or the detrimental consequences of poisoning. Health providers should ensure a correct understanding of treatment dosages and frequency intakes, as well as alarm signs regarding side effects or poisoning and how should parents and or patients act in a case of possible poisoning. They should also emphasize the hazard of having dangerous household products out of the reach of children. Further studies on the incidence of intentional injuries  in this population, and the effect of medication on the risk of poisoning are needed.Our findings have important implication from a public health standpoint. Poisoning remains a leading cause of preventable injuries in childhood and adolescence54, the Preferred Reporting Items for Systematic Review and Meta-Analysis (PRISMA)56 and the Meta-analysis of Observational Studies in Epidemiology (MOOSE)57 guidelines were followed when planning and carrying out our work. The protocol for the study was registered in the international prospective register of systematic reviews held by the University of York (PROSPERO) prior to data analysis (registration number CRD42017079911). Methods reflected those of our previous meta-analysis on the risk of physical injuries in order to make results comparable58.The Preferred Reporting Items for Systematic Review and Meta-Analysis Protocols (PRISMA-P)UNIKA: http://www.unav.edu/en/web/biblioteca), that uses the EBSCO discovery service (http://support.ebsco.com/help/index.php?lang=en&int=eds) to provide a list of references combining both internal (library) and external (database vendors) sources. Searches were carried out on November 30th 2017, with no time or language restrictions. References of retrieved pertinent papers were scanned to find additional possibly relevant studies. References of interest from our previous meta-analysis on the risk of physical injuries were also evaluated for potential inclusion17. See additional details, including search syntax, in Supplementary Methods Three major databases –PubMed (Medline Plus), Web of Science core database and Scopus- were searched. Furthermore, we searched over 110 additional databases from an institutional reference aggregator  or setting .59 and, similarly, poisoning injuries occur more frequently in males than in females1. Therefore, we included only articles that took into account this bias either by sample selection  or statistically (sex controlled as a confounding covariate).ADHD diagnosis is more common in males60, or equivalent constructs61, were excluded, since they are not equivalent to ADHD.The majority of the sample of a study had to be children and/or adolescents (defined as less than 18 years-old). The accepted operational definitions of ADHD were the following: (1) A categorical diagnosis according to standardized criteria, either the DSM  or the diagnosis of hyperkinetic disorder as per ICD-10 or previous versions; (2) Being above a pre-established threshold in a validated psychometric scale for the screening of ADHD; (3) The coding of the diagnosis in a medical registry; (4) A positive answer to the question: “Have you ever been told by a doctor that you have ADHD?” or similar questions; and (5) Being prescribed ADHD medication(s). Studies on preschoolers and those using the diagnosis of “Deficits in attention, motor control, and perception” (DAMP)The World Health Organization (WHO) definition of poisoning or intoxication was used to define eligible outcomes. Intoxication is defined by WHO as “a condition that follows the administration of a psychoactive substance and results in disturbances in the level of consciousness, cognition, perception, judgment, affect, or behavior, or other psychophysiological functions and response”1. Since the term “intoxication” can be used in relation to alcohol or drug abuse, outcomes describing poisoning were preferred to those reporting intoxications. However, if an article only reported intoxications, it was also included. Hence, outcomes from articles reporting health problems related to the codes T36-T61 of the 19th chapter of the International Classification of Diseases (ICD-10) or similar problems were deemed eligible.Poisoning cases had to be attended at medical settings and a registry created or self-reported. Studies reporting information requests to poison information centers or similar entities were not included. Poisoning could occur before or after the diagnosis of ADHD. Risk measures had to describe the ratio in the risk of poisoning between children and adolescents with and without ADHD. The primary outcome measure was defined as the hazard ratio (HR) obtained from Cox proportional hazards models, as it provides a time-independent estimation of the risk. However, odds ratios (ORs) are more frequently reported than HR, and they are the only risk measure that can be obtained from case-control studies that compare ADHD rates in a poisoned and a non-poisoned group. Therefore, ORs were accepted as secondary valid estimations of relative risk and combined with HRs.In order to address the secondary aim of the present meta-analysis , studies which reported outcomes on the relative risk of physical injuries in general and also provided similar data on the relative risk of poisoning were eligible.Studies were identified and selected following a two-stage process: (1) Two investigators independently and blindly screened retrieved titles and abstracts of all non-duplicated papers to exclude non-pertinent ones. Discrepancies were resolved by consensus; (2) Articles carried to this stage were assessed after reading the full-text following a similar process of double evaluation similar to the one of the previous step. Multiple reports of the same study were linked together.M.R.-G. and G.A. independently extracted data from articles that had been deemed eligible in the selection stage.A modified version of the Newcastle-Ottawa Scale (NOS)37 was used to assess the risk of bias of each study and rated independently by the same authors. Any discrepancies at this stage were resolved by consensus between M.R.-G. and G.A. Two items, one in the selection and one in the exposure subsections, were eliminated as they were deemed inadequate for our study. Hence, the maximum score in the scale was seven instead of nine stars. Final items of the scale can be found in Supplementary Methods 63.Summary effect measures (HRs or ORs) were combined in order to estimate a population-average relative risk between ADHD and poisoning in children and adolescents. Hazard Ratios and Odds Ratios were considered equivalent measures of relative risk as the baseline prevalence of poisoning was expected to be very low (less than 1%). In such cases the two measures yield very close results65, a statistical technique that models the nested structure between outcomes of the same study and allows to account for the non-independence of outcomes, was used for the inference of a mean effect size and meta-regression analyses. We carried out a mixed-effect model with robust variance and random-effect estimates. A model with variation of RRs between studies and equicorrelation between same-study effect sizes  was assumed. This strategy is highly efficient to estimate a mean model from outcomes which are typically correlated at the study level, but are usually independent between studies64. The influence of the equicorrelation value chosen here, the most commonly used in previous studies66, was evaluated in a sensitivity analysis with varying levels of p (0.1 steps between 0 and 1).Robust Variance Estimation2 index35 were used to evaluate heterogeneity among studies, whereas Begg’s adjusted rank correlation and Egger’s test were implemented to formally assess the presence of “small-sample” bias. These analyses were carried out using a single outcome per study. This outcome was selected at random whenever more than one existed. We planned to combine all outcomes fulfilling our inclusion criteria independently of the results of heterogeneity analyses and deal with heterogeneity through the use of a random-effects model and meta-regression, as the exclusion of studies prior to performing a meta-analysis affects the validity of the subsequent results67. Additional sensitivity/subgroup analyses consisted in: 1-Including ORs and HRs in separate analyses; 2-Calculating a mean effect size including only statistically unadjusted outcomes ; 3-Calculating a mean effect size including only adjusted outcomes ; 4-Evaluating the influence of removing articles that report intoxication risk (instead of poisoning); 5-Investigating the influence of the risk of bias as evaluated by the rating in the Newcastle-Ottawa Scale by carrying out a meta-regression analysis with the number of stars on the scale for each article as a predictor. We also investigated the effect of age on the risk of poisoning by splitting outcomes into 3 groups according to the age distribution of the participants: outcomes in which participants were under the age of 10, outcomes that included individuals of any age and outcomes in which participants were between 11 and 18 years old. The three groups of outcomes were compared pair-wise using a between-study meta-regression model as we hypothesized that the risk of poisoning would be significantly greater in the older group when compared to younger individuals or to individuals of all ages.Cochran’s Q test and I65. This analysis is optimal in cases where there exists within-studies variability in the covariate (outcome type in our case). This variability is studied by including the distance value around the study regressor mean as a covariate in the regression model68.Finally, we investigated whether, compared to those without, children/adolescents with ADHD had a significantly higher risk of poisoning than of suffering other kinds of physical injuries. To this end, in order to control for confounding variables which could affect this comparison , only studies that reported both outcomes  were evaluated. These two groups of outcomes were compared using a within-study meta-regression modelhttps://www.evidencepartners.com/resources/forest-plot-generator/).Effect sizes whose 95% confidence intervals (CIs) did not include 1 were considered significant. Analyses were carried out in STATA v13. Forest plots were created using the DistillerSR Forest Plot Generator from Evidence Partners (All data used in the preparation of the systematic review and meta-analysis is available upon request.Supplementary material


N = 163,136; 20,736 cases and 142,400 controls) by searching the PubMed and Google Scholar databases. Significant heterogeneity caused by different populations among the selected studies was observed. The A allele of rs965513 polymorphism was shown to be highly associated with risk of thyroid cancer, with odds ratios of 1.58 (95% CI 1.32–1.90) in all populations, 1.65 (95% CI 1.31–2.07)) in Caucasian populations and 1.49 in Asian populations. Compared to the dominant and recessive models, we observed the highest odds ratio  in the homozygous model. These results revealed that the rs965513 polymorphism is a risk factor for thyroid cancerPapillary thyroid cancer (PTC) is the most common type of thyroid cancer. With the rapid development of genome-wide association studies (GWAS), many genome variants associated with susceptibility to PTC have been identified, including the single nucleotide polymorphism rs965513 (9q22.33) near FOXE1. To evaluate the association between rs965513 and PTC in different ethnicities and countries, we conducted a meta-analysis using relatively large-scale samples from 23 studies ( Thyroid cancer (TC) is the most common malignancy in the endocrine system  and the P = 1.7 × 10−27) and rs944289 on 14q13.3  [The rs965513 polymorphism located near FOXE1 was first identified as significantly associated with thyroid cancer by Gudmundsson J in 2009. That study showed that two common variants are associated with thyroid cancer, specifically rs965513 on 9q22.33  .Meta-analysis has been described as combining and analyzing quantitative evidence from related studies to produce results based on a whole body of research . Thus, aForty-one articles were selected from the PubMed and Google Scholar databases. Based on the inclusion and exclusion criteria, 13 articles, including 23 independent studies, were included in our analysis. More detailed information about the decision to include or exclude the selected studies can be found in Figure vs. G) and the dominant model (AA + AG vs. GG), extreme heterogeneity was observed among the 23 selected studies . The recessive model (AA vs. AG + GG) and the homozygous model (AA vs. GG) showed large heterogeneity among the 23 selected studies . There was no significant heterogeneity observed in Asian populations. However, we found significant heterogeneity in Caucasian populations.The genetic heterogeneity of the rs965513 polymorphism was evaluated based on the additive, dominant, recessive and homozygous models and the data from the selected studies Table . SignifiWe used meta-regression to further investigate the potential sources of heterogeneity, including publication year, country, ethnicity, sample size, bias of sample size, type of control population and SNP genotyping techniques. We found that publication year, sample size and bias of sample size were significant, and these factors may be the sources of the observed heterogeneity . The absvs. 1.49), and this result could be strengthened by including more studies. Because 9 of the 23 selected studies provided the number of rs965513 genotypes or provided sufficient data to calculate the number of rs965513 genotypes, meta-analyses using dominant, recessive and homozygous models were conducted among these 9 studies. The association between rs965513 and thyroid cancer was also significant in the dominant model , the recessive model , and the homozygous model . In addition, the genotype AA was most strongly associated with risk of thyroid cancer in Caucasian populations , is a transcription factor involved in thyroid gland development  and vs. 142,400 controls) found in PubMed and Google Scholar to reevaluate this association, which will help to accurately assess the risk of rs965513 in PTC. Kang J. et al. included 8 studies corresponding to 2,085 cases and 10,341 controls [The significant association between the rs965513 polymorphism and thyroid cancer was first identified by Gudmundsson J in 2009 . Over thcontrols . Zhuang controls . Ai L etcontrols . Zhu et controls . Moreovecontrols . Ai L etcontrols . Zhu et controls .In our meta-analysis, the genetic heterogeneity of rs965513 among the selected studies was evaluated, and significant heterogeneity was observed in the additive, dominant, recessive and homozygous models. The genetic heterogeneity may be caused by differences in the publication year, sample size and bias of sample size through meta-regression. Because more studies corresponded to Caucasian populations, significant heterogeneity was observed in Caucasian populations but not in Asian populations. Next, meta-analyses were conducted separately for Caucasian, Asian and all populations. Our results showed that the A allele of rs965513 had a 1.58-fold risk of thyroid cancer in all populations, a 1.65-fold risk (95% CI 1.31–2.07)) in Caucasian populations and a 1.49-fold risk in Asian populations. Compared to the dominant and recessive models, the homozygous model showed the highest odds ratio  in all populations and in the Caucasian populations. The one-way sensitivity analysis suggested that the results of this meta-analysis were stable. We found that the Caucasian populations had higher risk than Asian populations, which further supports the previous findings.In addition to the rs965513 polymorphism, two other variants of FOXE1, rs1867277 and rs71369530, are also significantly associated with thyroid cancer . The varWe conducted meta-analyses using additive, dominant, and recessive models at the same time for more powerful results. However, our study has certain limitations. We attempted to obtain exact genotype numbers from all studies used in our analysis for the dominant and recessive models, but only some of the genotype numbers were available , 10–16. http://scholar.google.com/) to query the articles citing these studies and all of the references therein as identified using PubMed. We selected only published articles written in English.In stage 1, we searched PubMed and Google Scholar to select all possible studies with key words including ‘rs965513’ and ‘thyroid cancer’ or ‘FOXE1’ and ‘thyroid cancer’. The literature search was updated on April 7, 2015. Then, in stage 2, we used Google Scholar  the study provided sufficient data to calculate the OR and 95% CI.The studies that were selected met the following criteria: (1) the study must use a case-control design; (2) the study evaluated the association between rs965513 polymorphism and thyroid cancer; (3) the study provided the number of rs965513 genotypes or (4) provided sufficient data to calculate the number of rs965513 genotypes; and (5) the study provided an OR with 95% CI as well as the We extracted the following information from every study: (1) the name of the first author; (2) the year of publication; (3) the population or ethnicity; (4) the numbers of cases and controls; (5) the genotype number of rs965513 polymorphisms in cases and controls; (6) the number of rs965513 genotypes or (7) information needed to calculate the numbers of rs965513 genotypes; (8) the OR with 95% CI or (9) information needed to calculate the OR and 95% CI. All related calculations were completed using R software.The rs965513 polymorphism includes the two alleles G and A, of which A is the minor allele. A is assumed to be the high-risk allele and G the low-risk allele. We selected the additive, dominant, recessive and homozygous genetic model for further meta-analysis. The additive model can be described as the A allele versus the G allele .Q test was used to evaluate the genetic heterogeneity among the included studies, which approximately follows a χ2 distribution with k-1 degrees of freedom . I2 = (Q-(k-1)) ÷ Q × 100%, ranging from 0 to 100% [2 is a measure of heterogeneity and is a statistic that indicates the percentage of variance in a meta-analysis that is attributable to study heterogeneity [P < 0.01 and I2 > 50%. Meta-regression was used to investigate the potential sources of heterogeneity and was performed using the metafor package in R software.Cochran's  to 100% , was alsogeneity . Low, moogeneity . The sigFor case in which there is no significant heterogeneity among the included studies, the pooled OR was calculated using the fixed effect model; otherwise, the OR was calculated using random-effect model. The Z test was used to determine the significance of ORs. All statistical tests for heterogeneity and meta-analysis were performed using the meta package in R software.Funnel plots were used to evaluate the potential publication bias . Begg's


Therefore, we performed a systematic review and updated meta-analysis of the available evidence regarding the relationship between mTOR single nucleotide polymorphisms (SNPs) and cancer risk. Up to November 2015, 23 original publications were identified covering 20 mTOR SNPs, of which seven SNPs  were included in the final meta-analysis. We estimated the summary odds ratios (ORs) and corresponding 95% confidence intervals (CIs) for mTOR polymorphisms and cancer risk, and used the model-free approach to investigate the biological effect of each polymorphism. Our meta-analysis found that rs1883965, rs1034528, and rs17036508 were correlated with increased cancer risk in the complete over-dominant model . Stratifying analyses by cancer type, we found that the rs2295080 G allele was associated with a significantly higher risk of acute leukemia in the recessive model  and a lower risk of genitourinary cancers in the dominant model . Interestingly, further expression analysis showed that homozygous variant genotype carriers of rs1883965, rs1034528 and rs17036508 had lower mTOR transcript levels, based on HapMap data.mTOR regulates several cellular processes that are critical for tumorigenesis. However, previous studies on the association of  Deregularcinoma -10. Thesarcinoma , and sinarcinoma , have bemTOR, located on chromosome 1p36.2, is approximately 156 kb in length and is composed of 59 exons. According to the International HapMap Project Database (http://hapmap.ncbi.nlm.nih.gov/), 181 SNPs in mTOR have been reported in different populations, including those from Utah , China , Japan , and Nigeria . These 181 SNPs are distributed throughout mTOR, as well as 5kb upstream and downstream of the gene. Of these, 129 are considered common based on a minor allele frequency ≥ 5%, including seven SNPs in the 5′ upstream region, five SNPs in exonic regions, 112 in intronic regions, and five in the 3′ untranslated region (UTR) and downstream region.Human mTOR rs1057079 and colon cancer risk in 2010, clinical evidence has accumulated regarding the relationship between mTOR SNPs and the risk of various cancers, such as gastric cancer [mTOR and cancer risk. Since then, eight case-control studies  [mTOR SNPs and challenge conclusions from previous meta-analyses. The evidence is controversial for those mTOR SNPs investigated, partially because of insufficient statistical power. Consequently, we performed this updated meta-analysis to reassess the effect of mTOR polymorphisms within oncogenesis and to provide a more precise estimation of the associations.Since Slattery et al.  first rec cancer -18, esopc cancer , 20, endc cancer , renal cc cancer , acute lc cancer , 24, andc cancer . Previouc cancer  performestracts) , 27, 28 mTOR polymorphisms and meningioma [The selection process for eligible studies is shown in the flow diagram Figure . A totalningioma , which iningioma , 17, 18,ningioma , 24, twoningioma , 30, twoningioma , 20, andningioma , renal cningioma  and coloningioma . All stu2I = 62.4%, phet = 0.014). After excluding Li's study [P = 0.501), 0.97 (P = 0.504), and 1.20 (P = 0.422), respectively, suggesting a recessive effect of allele C. Therefore, the TC and TT genotypes were combined and compared with the CC genotype. A non-significant increase in cancer risk for the CC genotype was found .Seven studies, consisting of 6093 cases and 6442 controls, investigated the association between SNP rs2536 and cancer risk. We carried out a meta-analysis of rs2536 overall and in different cancer types under various genetic models. The seven studies were homogenous for OR1 and OR3, but heterogeneity was significant for OR2 , which discouraged us from calculating an overall estimate. Digestive system cancers  investigated in three studies were homogenous for OR1, OR2, and OR3, but still no significant association was observed. Accordingly, it appeared that rs2536 had no significant effect on susceptibility to cancer  were homogenous for OR1 and OR3, but heterogeneity was significant for OR2 , OR2  and OR3 . We failed to remove heterogeneity by excluding outliers identified by sensitivity analysis, so studies were grouped by cancer type to explore some possible major sources of heterogeneity , 0.77 (P < 0.001), and 0.99 (P = 0.955), respectively, suggesting a dominant effect of the G allele. Therefore compared with the TT genotype, carriers of the G allele (GG and TG genotypes) were shown to have a significantly reduced cancer risk   and OR2 . Consequently, there was no indication to pool the estimates. Childhood acute leukemia (acute lymphoblastic leukemia and acute myeloid leukemia), a non-solid cancer, was investigated in three studies, which appeared homogeneous and suggested a recessive model   were homogenous for OR1, OR2, and OR3, which were 0.76 (P = 0.733), 1.15 (P = 0.019), and 0.79 (P = 0.399), respectively. The summary estimate under the heterozygous model (GA versus GG) was statistically significant, implying that carriers of the rs1883965 GA heterozygote were more susceptible to cancer development. A comparison of GA versus GG/AA genotypes confirmed a complete over-dominant model , with the GA heterozygote having a higher cancer risk than either GG or AA homozygotes , so no further analysis was conducted.Three studies focused on digestive tract cancers . Heterogeneity was noted for OR1 (P = 0.791), 1.30 (P < 0.001), and 0.73 (P = 0.109), respectively, suggesting a complete over-dominant model . For rs17036508, heterogeneity tests were also negative for OR1, OR2, and OR3, at 0.99 (P = 0.975), 1.23 (P = 0.006), and 0.81 (P = 0.36), respectively, again suggesting a complete over-dominant model  , so it was not appropriate to meta-analyze the data. Two studies reporting rs1064261 genotype data from 1599 cancer patients and 1790 controls were homogenous for OR1, OR2, and OR3, with values of 0.90 (P = 0.82), 1.14 (P = 0.171), and 0.82 (P = 0.665), respectively. These estimates indicated a complete over-dominant model, whereas the overall gene effect was not significant   are located in transcription factor binding sites (TFBS), four  may affect exonic splicing enhancer or exonic splicing silencer binding site activity or even abolish a protein domain, and two (rs2536 and rs17036508) are within microRNA (miRNA) binding sites.Among the mTOR SNPs included in this meta-analysis, the F-SNP (FS) database found no functional information for the 3′-UTR SNP rs2536. Intronic polymorphisms rs3806317, rs17036508, and rs1883965 had an FS score of 0.101, while that of rs1034528 was 0.398, probably reflecting the frameshift coding changes it may cause. As a synonymous coding polymorphism, rs1064261 was shown to be conserved across multiple species with an FS score of 0.33. The mTOR promoter SNP rs2295080 was found to have an FS score of 0.101. Detailed information on the SNP functional bioinformatics analysis is shown in Table As for the mTOR transcript expression levels using the SNPexp web tool (Table P = 0.043)  and all populations . A similar trend was observed for rs17036508 for all populations , although no significant linkage disequilibrium was found between these two polymorphisms among the four different ethnicities  polymorphism in the i et al.  previousi et al. , rs2536 i et al.  focusingi et al. , gastrici et al. , prostati et al. , renal ci et al. , hepatoci et al. , and meni et al.  found noi et al. .The present pooled analysis also found no significant association between rs2536 and cancer risk, after removing the main source of heterogeneity and performing stratified analyses by cancer type. However, several studies indicated that although the main effect was not obvious, rs2536 was included in the best model to predict the risk of esophageal carcinoma  and prosmTOR upstream region was predicted to be within a TFBS by the SNPinfo database, which has been further confirmed by the lower nuclear protein binding activity of the G allele in human gastric cancer cell line SGC-7901 [mTOR mRNA levels compared with those with the wild-type T allele. Additionally, rs2295080 has been linked to decreased mTOR promoter activity in several cell lines [The rs2295080 (T > G) polymorphism located in the SGC-7901 . MoreoveSGC-7901  and renaSGC-7901  carryingll lines , 15, 25.ll lines , and renll lines , and thill lines . Howeverll lines , esophagll lines , and chill lines .We found that the rs2295080 G allele was associated with a significantly lower risk of genitourinary cancers in the dominant model, and a higher risk of acute leukemia in the recessive model. Therefore, the biological effect of rs2295080 might be cancer-specific. Notably, an obvious divergence of rs2295080 effects was observed in cancers of the digestive system, which might be partially explained by a high degree of heterogeneity, especially for gastric cancer [mTOR, so is more likely to be involved in the regulation of transcription and be associated with disease compared with SNPs in other introns [mTOR mRNA expression in YRI individuals carrying the rs1883965 heterozygote GA (P = 0.043). Such discrepant results may not be entirely attributed to racial differences, but could reflect the small sample size of gene expression data, which increases the probability of false-positive findings. Therefore, more studies are essential to obtain a more reliable conclusion regarding the association between rs1883965 and mTOR transcription.The SNP rs1883965 (G > A) is located within the first intron of  introns . The SNP introns  and gast introns , and ourmTOR. Bioinformatics analysis revealed that rs1034528 is located within a TFBS and causes a frameshift coding change. Its FS score of 0.398 is the highest among those mTOR SNPs investigated. SNP rs17036508 was predicted to be located within a miRNA binding site and an exonic splicing enhancer or silencer motif, affecting the splicing of pre-RNA. Previously, the associations between rs1034528 and rs17036508 and gastric and prostate cancer were investigated and the rs1034528 C allele was shown to be a risk factor in two independent studies [mTOR transcript expression levels. Because the mTOR signaling pathway usually promotes oncogenesis, this finding is unexpected but could be explained by the fact that mTOR mutants caused by different amino acid substitutions have different abilities to phosphorylate substrates S6K1 and 4E-BP1, even though they are expressed at similar levels after nutrient starvation [ANGPTL7), itself within intron 28 of mTOR. ANGPTL7 expression was reported to be up-regulated by hypoxia in cancer cells and to exert a pro-angiogenetic effect, which is essential in the early stages of tumor development [mTOR and ANGPTL7 simultaneously.SNPs rs1034528 (G > C) and rs17036508 (T > C) are both located within intronic regions of  studies , 30. In arvation . Thus, irs1064261 (T > C) and rs1057079 (A > G) are synonymous SNPs within exonic regions. Although not altering amino acid sequences, they were predicted to interrupt the exonic splicing enhancer or silencer motif, or even abolish a protein domain. Positive correlations of the rs1064261 C allele with an increased risk of neuroendocrine tumors  and gastAlthough several intronic SNPs were excluded from the present meta-analysis and the integrated results of their relationships with cancer risk were not discussed because of the limited number of studies, some of them exhibited potential biological activity. SNPs rs12125777 (C > T) and rs12124983 (C > T) were identified to interfere with transcriptional regulation by bioinformatic analysis; these predictions were confirmed by clinical evidence , 53 and mTOR SNPs studied to date only have a mild effect on cancer development. The FS integrative scoring system also defined these polymorphisms as having relatively moderate deleterious effects, because no mTOR SNPs have yet been assigned an FS score as high as 0.5, which is the median score of disease-related SNPs [Integrated risk estimates of clinical evidence reveal that the ted SNPs .mTOR SNP on cancer susceptibility appears to be weak, its interaction with functionally relevant variants and environmental factors might have a greater effect on oncogenesis. Carriers of combined risk alleles have been previously shown to have a significantly increased risk of developing various types of cancer, mostly in a dose-dependent manner. For example, Li et al. [mTOR polymorphisms  exhibited a higher susceptibility of developing prostate cancer  compared with those with one or zero adverse genotypes. Such cumulative effects were also observed across SNPs in mTOR and mTOR pathway-related genes  [mLST8 and RPTOR) [mTOR haplotype effects on cancer risk [Although the magnitude of the effect of an individual i et al.  found thd RPTOR) . Severalcer risk , 49. ZhumTOR polymorphisms and environmental factors such as smoking status, drinking status, age, sex and BMI have also been reported. Both the effects of individual SNPs and combined risk genotypes were greater in some high-risk subgroups for many cancers such as older adults and smokers [mTOR SNPs and BMI were verified in esophageal squamous cell carcinoma [mTOR variants and cancer susceptibility. However, the underlying mechanisms may vary depending on cancer type and remain to be investigated.Possible interactions between  smokers , 29, 30.arcinoma  and breaarcinoma . It has arcinoma . BecausemTOR SNPs and cancer risk, the pooled sample size of this meta-analysis was still relatively small, especially for rs1034528, rs17036508, rs1064261, and rs3806317. This weakened the statistical power and limited our ability to perform more accurate subgroup analyses for specific cancer types. Second, the ethnicity of all available studies for meta-analysis was Han Chinese, so our findings may not be applicable to other populations. However, because Han Chinese is the largest ethnic group in the world and as the gene pool reflects a long history of immigrations and intermarriages with other ethnic groups [mTOR polymorphisms on cancer development. Third, all studies included in the systematic review were published in English or Chinese, yet publications in other languages may contain different relevant studies. This may be the main source of publication bias in our meta-analysis. Finally, although the SNPinfo database and F-SNP aim to reduce the number of false-positive results, computational predictions of certain SNPs are only estimates and should be confirmed by functional studies.A number of limitations of this systematic review and meta-analysis should be noted. First, although we collected all published clinical evidence investigating c groups , these dmTOR polymorphisms has mainly focused on cancer susceptibility in recent years. To date, only 20 of 129 common SNPs within mTOR have been investigated in relation to cancer risk. Future studies may benefit from genotyping additional polymorphisms to identify more functionally significant variants. Additionally, although the risk effects of the mTOR SNPs so far examined are too small to be regarded as clinically useful, their interactions with other genetic variants or environmental factors have been shown to contribute to further increases in cancer risk either additively or synergistically. The mechanisms of these joint effects deserve further research. mTOR polymorphisms may also be associated with clinical outcomes and response to chemoradiotherapy. Several polymorphisms, such as rs2295080, rs11121704, and rs12139042, have been shown to be significantly associated with lung and esophageal cancer [mTOR polymorphisms with treatment response to inhibitors of the mTOR pathway. In vitro functional studies should also be conducted to confirm these functional predictions and reveal the underlying molecular mechanisms behind the observed associations. Finally, some investigations have found that mTOR inhibitors are able to prolong lifespan not just by inhibiting the growth of tumors, but by postponing the aging process [mTOR polymorphisms may also play roles in other age-related diseases, such as cardiovascular and neurodegenerative disorders, and further investigation is required.The study of l cancer -60, but  process , 61. ThiThis systematic review and meta-analysis was conducted following the PRISMA  recommendations .We searched the PubMed, Embase, Chinese National Knowledge Infrastructure and Wanfang Data databases to identify potentially relevant studies published before November 3, 2015, without language restrictions. The following keywords were used jointly as search terms: “mTOR” or “FRAP” or “RAFT1” or “RAPT1”, “polymorphism” or “variant” or “mutation” and “cancer” or “tumor” or “carcinoma” or “carcinoma” or “malignancy”. The full electronic search strategy for PubMed is shown in Appendix 2. We also manually searched the reference lists to identify other potential articles. If overlapping data by the same first author were found, the article with the largest number of subjects was included.mTOR polymorphisms and cancer risk; (2) written in English or Chinese; and (3) case-control studies. Studies were excluded if they were: (1) reviews or comments; or (2) animal or cell line studies. Eligible studies were determined by two researchers (ZNJ and LX) independently. Disagreement was resolved by discussion or consulting another researcher (CYH).Studies were eligible if they met the following criteria: (1) evaluated the association between Relevant information, including the first author's name, year of publication, country in which the study was conducted, ethnicity, cancer type, control source , genotyping methods, matching criteria for controls, number of cases and controls, and genotype distribution of cases and controls, was extracted from each eligible study by two independent researchers (ZNJ and LX).The quality of eligible studies in this analysis was evaluated according to the NOS , which cP < 0.05 was considered statistically significant.HWE was evaluated in the controls of each study using the chi-square test. The crude odds ratio (OR) and corresponding 95% confidence interval (CI) were calculated by the Z test to assess the strength of the association between genotype and cancer risk; mTOR polymorphisms in the occurrence of malignancies, we used the methods recommended by Thakkinstian [versus WW, WV versus WW, and VV versus WV for each polymorphism that qualified for meta-analysis to detect the existence of heterogeneity. Appropriate genetic models were then determined in terms of the relationship between the three pairwise differences:To identify the best matching genetic model for kinstian . OR1, OR(1) Recessive model: if OR1 = OR3≠1 and OR2 = 1.(2) Dominant model: if OR1 = OR2≠1 and OR3 = 1.(3) Complete over-dominant model: if OR1 = 1, OR2 = 1/OR3≠ 1.(4) Co-dominant model: if OR1 > OR2 > 1 and OR1 > OR3 > 1, or OR1 < OR2 < 1 and OR1 < OR3 < 1.Using the indicated genetic model collapsed the three genotypes into two groups (except in the case of a co-dominant model):versus WW.(1) If a dominant model was indicated, V carriers (VV plus WV) versus W carriers (WV plus WW).(2) If a recessive model was indicated, VV versus WV.(3) If a complete over-dominant model was indicated, (VV plus WW) versus WV, and VV versus WW.(4) If a co-dominant model was indicated, VV 2I test [P < 0.1 or 2I > 50 %) and homogenous studies, respectively. Potential sources of heterogeneity were explored using sensitivity analysis carried out by removing a single study from the meta-analysis each time, or subgroup analyses with cancer types. Potential publication bias was estimated using Begg's rank correlation [P < 0.1 was considered significant). STATA software version 11.0  was used for statistical analyses.Between-study heterogeneity was evaluated using the Q-statistical test and I2 test . The ranI2 test  and fixeI2 test  were takrelation  and Eggerelation  (P < 0.1http://snpinfo.niehs.nih.gov/snpinfo/snpfunc.htm) and the functional single nucleotide polymorphism (F-SNP) database [http://compbio.cs.queensu.ca/F-SNP/). Potential biological effects of the investigated mTOR SNPs were evaluated, including changes in protein coding, transcriptional regulation, splicing sites, and micro (mi)RNA-binding sites. Correlations between mTOR polymorphism genotypes and gene expression levels from 270 HapMap phase II and III individuals from four populations  were conducted online, using the SNPexp web tool [http://tinyurl.com/snpexp).To cover as many bioinformatics web services and public databases as possible, we used two integration platforms of SNP analysis resources: the SNPinfo database  (http://database , 71 (httweb tool  (http://


N‐acetylcysteine (NAC) pretreatment in the prevention of contrast‐induced nephropathy (CIN). In this meta‐analysis of randomized controlled trials, we aimed to assess the effectiveness of NAC treatment for the prevention of CIN.Conflicting results have been obtained in trials that have evaluated the prophylactic efficacy of EMBASE, and the Cochrane Library were electronically searched from inception to January 2016 for all relevant studies. The weighted relative risk (RR) and corresponding 95% CI for incident CIN were estimated using random effects models. Standard methods for assessing statistical heterogeneity and publication bias were used. The study included 11 480 participants and 1653 cases of CIN. The incidence of CIN was 12.8% in the NAC group versus 16.0% in the control group . In the patients undergoing coronary angiography, the incidence of CIN in the NAC group versus the control group was 13.7% versus 17.2% ; in those undergoing peripheral angiography, the incidence was 6.4% versus 5.8% ; in those undergoing computed tomography, the incidence was 7.7% versus 14.8% .PubMed, NAC supplementation and risk of CIN in patients undergoing coronary angiography and computed tomography, while a protective role for NAC in patients undergoing peripheral angiography was not obvious.Our meta‐analysis showed an inverse and significant association between  Contrast‐induced nephropathy (CIN) is a quite common and well‐known complication following the administration of iodinated contrast media and has become the third most common cause of hospital‐acquired acute kidney injury after hypotension and surgery.The precise mechanism leading to CIN has not been fully elucidated. There is evidence that contrast agents reduce renal function through a combination of renal vasoconstriction with consequent hypoxia, and direct toxicity on tubular epithelial cells.N‐Acetylcysteine (NAC) is a thiol‐containing, cell‐membrane‐permeable antioxidant. The benefit of NAC supplementation for the prevention of CIN in patients with renal insufficiency undergoing contrast‐enhanced computed tomography (CT) was first reported by Tepel et alThere are several possible mechanisms underlying the association between NAC administration and CIN risk. NAC has the potential to prevent CIN risk due to its potent antioxidantThere have been few approved therapies for CIN. The current standard of care involves only the use of intravenous hydration and low‐osmolality contrast media, but the benefit of this approach is limited.N‐acetylcysteine or NAC or acetylcysteine) and (contrast media or contrast agent or contrast‐induced nephropathy or contrast‐associated nephropathy or radiocontrast nephropathy or contrast nephrotoxicity or acute kidney failure or acute kidney injury). We further restricted the search to studies on humans and those written in English. Additional studies not captured by our database search were retrieved through a manual search of references from originally identified reviews and research reports. This process was repeated until no additional articles were identified. Because this was an analysis of previously published data, this study did not undergo or require Institutional Review Board approval.Relevant studies were identified by searching PubMed, EMBASE, and Cochrane Library databases from their inception to January 2016 to identify the association between NAC supplementation and CIN risk using the following search terms:  randomized controlled trials involving adult patients undergoing coronary angiography or peripheral angiography or CT that assessed the anti‐CIN efficacy of NAC supplementation; (2) use of NAC as monotherapy or only in combination with hydration, with a control group that received placebo or hydration; (3) definition of CIN as an absolute increase in serum creatinine of ≥0.5 mg/dL (44 mmol/L) or a relative increase of ≥25% from the baseline value after the administration of contrast media; and (4) studies published in English. Exclusion criteria were as follows: (1) patients treated with both NAC and other drugs (except hydration); (2) trials with abstracts only; and (3) patients undergoing renal replacement or those with coexisting cancer or malignant disease.Two investigators (X.R.F. and C.G.Z.) independently reviewed all relevant articles and identified eligible studies. Disagreements or uncertainties were resolved by consensus. The following data were extracted from each study: first author's name, publication year, geographic region, sample size, subject characteristics , definition of CIN, dosage of NAC and contrast agent, administration route , and the intervention in the control group. The primary outcome was the development of CIN, defined as an absolute increase in the serum creatinine concentration of at least 0.5 mg/dL or a >25% from the baseline value that occurred within 2 to 5 days after contrast injection. In case of trials in which the incidence was reported in terms of both relative (by 25%) and absolute increase in creatinine (by 0.5 mg/dL) separately, the data for the relative increase were given preference on the basis of advantages of this approach.The quality of the studies was assessed through the methods used by Moher et al.P values, or t or F statistics according to standard formulas.2), where a P value <0.10 or I2>50% was considered to be significant.P value <0.05 was considered to be statistically significant in this trial, unless otherwise specified. Publication bias was assessed by using funnel plots, Begg's test, and Egger's test.Statistical analysis was performed using the Review Manager 5.0 (Cochrane Collaboration) and STATA software version 12.0 (Stata Corporation). For dichotomous outcomes, the results were expressed as risk ratios (RRs) with 95% CIs. For continuous outcomes, the results were expressed as weighted mean difference with SD. For trials that did not report SDs, SD values were obtained from 95% CI, P value <0.05 was considered to be statistically significant in this trial unless otherwise specified.In addition, to further detect and evaluate clinically significant heterogeneity, subgroup analyses and univariate meta–regression analyses were conducted to explore potential effect modification by prespecified factors: different procedure method, NAC dosage, NAC administration route, baseline renal function, contrast agent dosage, Jadad score, and CIN definition. A We initially retrieved 2703 potentially relevant articles from the database, and 2467 articles were determined to be irrelevant after screening of the title or abstract. We conducted a detailed evaluation of the complete report for 236 trials proceeded to a detailed evaluation of the complete report, following which a further 175 articles were excluded. Finally, the 61 remaining articles were included in our meta‐analysis.The 61 articles with 66 comparisons were published between 1996 and 2016 and yielded a cumulative total of 11 480 patients, 5757 of whom were randomly assigned to the NAC group and 5723 to the control group (Table P=0.0002), while the heterogeneity was significant  . The incidence of CIN was 12.8% (739 of 5757) in the NAC group and 16.0% (914 of 5723) in the control group; in the pooled analysis using a random effects model, patients receiving NAC had a 24% lower risk of CIN than the control group , with significant heterogeneity . Low heterogeneity was seen with this analysis  .There was no evidence of heterogeneity  , with significant heterogeneity , with significant heterogeneity . Twenty comparisons with 3307 patients reported rates of CIN risk in patients with high doses of contrast agent , with significant heterogeneity . Fourteen comparisons with 2335 patients reported an association between NAC intake and CIN risk in patients with diabetes, although the results were not significant . There was no evidence of heterogeneity    .The meta‐regression indicated that the impact of NAC on risk of CIN was consistent over baseline renal function P=0.55 FigurP=0.55  with significant heterogeneity  . The incidence of nephropathy requiring dialysis was extremely low, and only 43 cases were reported among 7168 randomized patients . The overall results indicate that NAC admission does not significantly reduce the incidence of renal failure requiring renal dialysis . Mortality within 30 days occurred in 213 of the 6973 randomized patients, 2.8% in the NAC group versus 3.3% in the control group  .Compared with the control group, a significant reduction in blood creatinine level was observed in the NAC group  and Egger's test (P=0.44) provided no evidence of publication bias.The quality of these 61 randomized controlled trials was variable. Thirty studies with 33 comparisons were classified as high quality (Jadad score of 4 or 5), and the other 31 studies with 33 comparisons were classified as low quality (Jadad score of 2 or 3) (Table S1). The funnel plots of the studies were symmetric by visual in the current meta‐analysis . In addition, the Begg's test  and found that NAC intake has a sufficient effect on inhibition of CIN risk on both groups. Furthermore, a higher volume of contrast agent is more frequently needed if endoluminal therapy is required, and this is also a risk factor for CIN. In addition, subjects undergoing coronary angiography and CT, instead of peripheral angiography, may experience the maximum benefit of NAC on CIN inhibition. It is worth considering potential difference related to CIN resulting from different procedures. Patients undergoing coronary angiography are likely to have some baseline diseases, such as coronary disease, diabetes, or renal dysfunction, which are also CIN risk factors. However, only limited studies have analyzed the effect of NAC intake on CIN risk in patients undergoing peripheral angiography (5 studies) or CT (7 studies); therefore, more randomized controlled clinical trials will be required to make a definite conclusion.The analysis of our secondary outcomes revealed a significant improvement in the blood creatinine level with NAC supplementation, which was consistent with our findings that CIN risk was significantly decreased. However, this analysis did not support the use of NAC to reduce the incidence of CIN in patients with diabetes, or nephropathy requiring dialysis. Diabetes mellitus has been regarded as an important risk factor for CIN. However, as only 2335 patients with diabetes, with 351 cases of CIN, were enrolled in this meta‐analysis, we could only observe a tendency instead of an obvious inhibition of CIN risk. In addition, NAC intake demonstrates only a tendency instead of significant protection from mortality.This meta‐analysis has several significant strengths. First, to our knowledge, this study represents the largest available pooled analysis to date evaluating NAC efficacy for CIN prevention. The populations studied varied widely and covered several major risk factors for CIN, which enabled us to draw clinically relevant conclusions from different subsets of populations. Second, the trials included in this study were all randomized controlled trials, with careful monitoring and adjudication by blinded clinical events committees, which ensured the relatively high quality and the accurate information of the included studies. Finally, since the definition of CIN varied across studies, we chose change in the blood creatinine level rather than acute renal failure or requirement for dialysis as our primary outcome; thus, the differential misclassification of CIN attributable to recall bias was minimized.Our analyses did have limitations. First, the sample sizes in most of these trials were relatively small, with numbers of patients less than 200; thus, meta‐analysis may have been underpowered to detect true differences. Second, a significant amount of unexplainable heterogeneity was detected in both primary and subgroup analyses, although our random effects model did account for this heterogeneity. It is possible that the baseline characteristics of the participants all contribute to variation in trial effects. Although the variables of contrast agent dosage account for part of the statistic heterogeneity, the residual heterogeneity remained at 40%. Third, we did not have access to patient‐level data to determine whether preexisting decreased renal function and other risk factors  could influence the effect of NAC intake on CIN risk. Fourth, the follow‐up period for most included studies was only 48 or 72 hours. CIN can occur beyond 2 days, peaking on the fifth day. Therefore, some patients developing CIN beyond 48 hours have been missed. Fifth, an obvious source of conflict was that there is no general agreement on the safe dosage of NAC. In the trials of this meta‐analysis, NAC dosage ranged from 600–7200 mg/days; therefore, it is difficult to determine the optimal dose that would lead to the greatest improvement in renal function with limited side effects. Sixth, there is publication bias between studies, which questions the reliability of the results.In conclusion, this meta‐analysis provides strong evidence that NAC supplementation is associated with a significantly lower risk of CIN. In real‐world practice, it is impossible to provide NAC for all patients undergoing contrast agent injection, while it is reasonable to administer NAC by the oral route for patients who are undergoing coronary angiography and who have renal dysfunction or who are receiving high doses of contrast agent. Additional randomized controlled trials with longer terms and larger populations are required to establish causality and to elucidate the underlying mechanisms.R.‐F.X. and G.‐Z.C. conceived the study design, and wrote the manuscript; A.‐Y.T., Y.B., and Y.‐B.D. performed the analyses. All authors read and approved the final manuscript.The present study was supported by the National Natural Science Foundation of China (Nos. 81400369 and 81500293).None.Table S1. Quality assessment of included studies.Figure S1. Meta‐analysis of effects for NAC (N‐acetylcysteine) on serum creatinine compared with control arms. IV indicates intravenous.Figure S2. A, The association between NAC (N‐acetylcysteine) admission and the incidence of renal failure requiring renal dialysis. B, The association between NAC admission and the incidence of mortality. RR indicates risk ratio.Figure S3. Funnel plot of N‐acetylcysteine consumption and contrast‐induced nephropathy. The SE of the risk ratio (RR) was plotted against the RR for contrast‐induced nephropathy.Click here for additional data file.