Aflibercept is a human recombinant fusion protein with antiangiogenic effects that functions as a decoy receptor to bind vascular endothelial growth factor A. Proteinuria is one of its major adverse effects with a substantial variation in the incidence rate, and the overall risk of proteinuria has not been systematically studied. We performed a meta-analysis of published clinical trials to quantify the incidence and relative risk of proteinuria in cancer patients treated with aflibercept.The electronic databases were searched, including PubMed, Embase, Cochrane databases, and ASCO  abstracts. Eligible studies were phase II and III prospective clinical trials of cancer patients treated with aflibercept with toxicity data on proteinuria. Overall incidence rates, relative risk (RR), and 95% confidence intervals (CI) were calculated using fixed or random effects models depending on the heterogeneity of the included studies.A total of 4,596 patients with a variety of solid tumors from 16 prospective clinical trials were included for the meta-analysis. The overall incidences of all-grade and high-grade proteinuria in cancer patients were 33.9% (95% CI: 27.3–42.1%) and 7.9% (95% CI: 6.1–10.2%). The relative risks of proteinuria of aflibercept compared to control were increased for all-grade  and high-grade  proteinuria. The risk of developing all-grade and high-grade proteinuria with aflibercept was substantially higher than that of bevacizumab .Aflibercept is associated with an increased risk of developing proteinuria. Appropriate monitoring and treatment is strongly recommended to prevent potential renal damage. Future studies are still needed to investigate the risk reduction and possible use of aflibercept in cancer patients. Angiogenesis is the formation of new blood vessels, which is an important process in the growth of malignant tumors. The predominant regulator of tumor angiogenesis is vascular endothelial growth factor (VEGF) Aflibercept (Ziv-aflibercept), also know as VEGF Trap, is a recombinant fusion protein comprised of the extracellular domain from VEGFR-1 and VEGFR-2 fused with Fc region of human IgG1. It is a circulating antagonist that binds to VEGF-A, VEGF-B and PIGF , subsequently preventing their interaction with VEGFR-1 and VEGFR-2, which is a more potent VEGF blocker than bevacizumab Although aflibercept appears to be well tolerated, as with other anti-angiogenic inhibitor, aflibercept may cause some adverse effects. Asymptomatic proteinuria is common in patients treated with anti-VEGF inhibitors. The recognition and management of proteinuria in cancer patients treated with aflibercept is an important issue since proteinuria may be related with renal damage. The risk factors are not well understood. Because of the limited number of patients in each trial, the overall risk of proteinuria with aflibercept is unclear. Thus, we performed a meta-analysis of prospective clinical trials to determine the incidence and relative risk of proteinuria among cancer patients treated with aflibercept.The electronic databases were searched for studies to include in the meta-analysis, including PubMed, Embase, and Cochrane databases. Abstracts presented at the annual meetings of the American Society of Clinical Oncology (ASCO) were also searched manually. The upper date limit of March 2014 was applied, with no lower date limit. Searches include the terms:  And , And . The references cited by the included studies were also used to complete the search.Aflibercept had been approved for the treatment of patients with previously treated colorectal cancer at a recommended dose of 4 mg/kg every 2 weeks (Q2W). Trials using aflibercept at the approved dosage were included. Clinical trials using aflibercept at doses of 6 mg/kg every 3 weeks (Q3W) were also included to assess the possible increased incidence of proteinuria with these treatments.Eligible criteria for inclusion in this meta-analysis are: (1) prospective phase II and III clinical trails in cancer patients; (2) participants assigned to treatment with single agent aflibercept at 4 mg/kg Q2W or 6 mg/kg Q3W; (3) the language was restricted in English; (4) data available regarding events or incidence of proteinuria, and (5) if multiple publications of the same trial were retrieved, only the most recent publication (and the most informative) was included. Phase I studies were excluded because of the different drug dosage and the relatively small number of patients on these trials. Abstracts of all candidate articles were read by two independent readers (LP and YZ). Articles that could not be categorized based on title and abstract alone were retrieved for full-text review. Disagreements were resolved by consensus between the two readers. To determine the issue of multiple publications from the same data sets, we checked all author names, clinical trial information, and the time period of patient recruitment of the articles.http://ctep.cancer.gov/reporting/ctc_archive.html). The CTC version 3.0 describes the grading of proteinuria as follows: grade 1, 1+ or 0.15–1.0 g/24 hrs; grade 2, 2+ to 3+ or >1.0–3.5 g/24 hrs; grade 3, 4+ or >3.5 g/24 hrs; and grade 4, nephrotic syndrome. We included all incidences of proteinuria of grade 1 or above in our analysis.Two investigators independently assessed the eligibility of the articles and abstracts identified by the search, and discrepancies were resolved by consensus. Proteinuria was extracted from the safety and toxicity profile in the primary study. These clinical end points were all recorded according to versions 3.0 of the Common Terminology Criteria for Adverse Events (CTCAE) of National Cancer Institute (Two authors (LP and YZ) independently assessed the risk of bias in the 5 included randomized controlled trials using RevMan 5.3. Agreements were reached by discussion between the two review authors if there were disagreements on specific items in the studies.2-based Q statistic P<0.05 or I2>50%. If heterogeneity existed, data were analyzed using a random effects model. In the absence of heterogeneity, a fixed effects model was used. To calculate the pooled incidence, an inverse variance statistical method was used. A statistical test with a P value less than 0.05 was considered significant. To assess the stability of results, sensitivity analysis was carried out by sequential omission of individual studies. To test for variation in incidence estimates by other factors, we conducted a meta-regression analysis. The presence of publication bias was evaluated by using the Begg's and Egger's tests Information was retrieved from the primary studies, using a standardized data collection form, including the following items: year of publication, first author, underlying malignancies, number of patients, treatment arm. If data from any of the above categories were not reported in the study, items were treated as “NR (not reported)”. The data of the number of patients with all-grade and high-grade (grade 3 and grade 4) of proteinuria and the number of patients receiving single agent aflibercept were extracted from the toxicity profile. For each study, we derived the proportion and 95% confidence interval (CI) of patients with proteinuria. For studies with a control arm in the same trial, we also calculated and compared the relative risk (RR) of proteinuria. For one study that reported zero events in the control arm, we applied the classic half-integer correction to calculate the RR and variance Our search yielded a total of 256 articles on aflibercept from the literature. After reviewing each publication, 15 original studies of full publication met our inclusion criteria. From the abstracts published in American Society of Clinical Oncology (ASCO) meetings, 1 abstracts related to aflibercept were also identified. Altogether, 16 primary studies met inclusion criteria in the search strategy and study selection section, comprising 4596 patients for final analysis . The majI2 = 98.8%, P = 0.00), and the calculated summary incidence of all-grade proteinuria among patients receiving aflibercept was 33.9% (95% CI: 27.3–42.1%) using a random effects model (P>0.2).The results of the meta-analysis were shown in ts model . We atteet al in patients with melanoma I2 = 82.4%, P = 0.00)  .P = 0.002, I2 = 80.4%, P = 0.00). The incidence for high-grade proteinuria was significantly increased in cancer patients receiving aflibercept compared with control .With a view to investigate the specific contribution of aflibercept to the development of proteinuria and exclude other therapeutic interventions, we then determined the relative risk (RR) of aflibercept–induced proteinuria compared with control arm. The pooled RR showed that aflibercept treatment increased the risk of developing all-grade proteinuria in cancer patients with a RR of 1.41  initiates angiogenesis signaling process, including increased vascular permeability and endothelial cell proliferation Aflibercept  is a recombinant protein consisting of domain 2 from VEGFR-1 fused to domain 3 from VEGFR-2, attached to the hinge region of the Fc domain of IgG1. In contrast to bevacizumab, aflibercept not only targets VEGF-A, but also VEGF-B and PIGF, forming a pharmacologic blockade of the VEGF pathway. Aflibercept has a higher VEGF A binding affinity than bevacizumab Proteinuria is one of the major side effects of this drug, and reported incidences vary substantially among clinical trials. The underlying mechanism is not entirely understood. VEGF plays an important role in regulating glomerular vascular permeability. Treatment of mice with a single dose of anti-VEGF agent resulted in proteinuria The aim of this study is to gain a better understanding of the overall incidence and relative risk of proteinuria in patients with cancer who receive aflibercept. The present meta-analysis has combined 16 publications including 5 randomized controlled trials and 11 phase II trials. Our meta-analysis results demonstrate that aflibercept is associated with an increased risk of developing proteinuria. The overall incidence of all-grade and high-grade proteinuria was 33.9% (95% CI: 27.3–42.1%) and 7.9% (95% CI: 6.1–10.2%), respectively. The relative risks of proteinuria of aflibercept compared to control were increased for all-grade  and for high-grade  proteinuria. Data were insufficient to analyze the differences of various underlying malignancies.We also explore the difference in the incidence of proteinuria associated with aflibercept compared with bevacizumab. The results show that the risk of developing proteinuria with aflibercept is substantially higher than that of bevacizumab. Aflibercept and bevacizumab have different blocking site of the angiogenic pathway. It is possible that the blockade of VEGFR rather than VEGF would result in different downstream effects and toxicities. Since there was no clinical trials which directly compared aflibercept and bevacizumab, the results should be explained with caution. As the development of aflibercept continues, this agent will come to head-to-head comparison with bevacizumab and VEGFR TKIs .Our meta-analysis demonstrates that proteinuria associated with aflibercept is mostly grade 1 and 2. The drug manufacturer recommends monitoring for proteinuria by urine dipstick  and determination of the urinary protein-to-creatinine ration (UPCR) prior to each dose of aflibercept Our meta-analysis has several limitations. One limitation of our meta-analysis is that these studies are conducted at various institutions by different investigators and may have potential bias in reporting the types of adverse events. Secondly, our meta-analysis was based on data from trials that have published results in the literature, but not individual patient data In summary, our meta-analysis is the first study to systematically estimate the incidence and relative risk of proteinuria associated with aflibercept in cancer patients. The current analysis suggested that the use of aflibercept increased the risk of all-grade and high-grade proteinuria. The relative risks of proteinuria of aflibercept compared to control were increased for all-grade and high-grade proteinuria. These results would provide important information for clinicians who use aflibercept to treat patients with solid tumors.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


TSHR rs12101255 and rs2268458 polymorphisms had no association between GO and GD (GD without GO). The results indicate that rs179247 and rs12101255 are likely to be genetic biomarkers for GD. Further studies with different population groups and larger sample sizes are needed to confirm the genetic associations of the TSHR gene with GD/GO.Graves’ disease (GD) is a common thyroid disease, and Graves ophthalmopathy(GO) is the most common extra-thyroidal manifestation of GD. Genetic associations of the thyroid stimulating hormone receptor (TSHR) gene with GD and GO have been studied in different population groups for a long time. We aimed to obtain a more precise estimation of the effects of TSHR single nucleotide polymorphisms (SNPs) on GD/GO using a meta-analysis. Publications were searched on Pub Med and EMBASE up to December 30, 2015. Eight studies involving three SNPs , which included 4790 cases and 5350 controls, met the selection criteria. The pooled odds ratios (OR) and the 95% confidence intervals (CI) were estimated. SNPs rs179247 (dominant model [GG + GA vs. AA]: OR = 0.66, 95%CI: 0.61–0.73, P = 0.000, I Graves’ disease (GD) is the most important factor of hyperthyroidism and is a typical autoimmune thyroid disease3Thyroid stimulating hormone receptor (TSHR) is expressed by the plasma membrane of thyroid follicular cells (TFC)The pathogenesis of GD/GO is complex and multifactorial with several environmental and genetic factors. Because TSHR is the main auto antigen in GD and because the titre and the activity of the TSHR autoantibody are positively associated with the severity of GOGenetic associations of immunoregulation and thyroid-specific genes with GD/GO have been studied in different people for a long time. At first, three SNPs  were studied. However, they were finally excluded because of their frequent presence in healthy people1416First, 754 articles were selected from the databases, and 252 duplicate articles were removed. Among the remaining articles, 459 were not related to the topic. A total of 43 relevant articles were identified for further study. Of those, 35 articles did not meet the inclusion criteria, including 6 reviews. Additionally, 1 study did not have sufficient data, and 28 studies did not study the three SNPs. In the end, we selected 8 studies that met the inclusion criteria, including 5 studies that involved 4821 GD patients and 4846 healthy controls121920212022232421242 = 58.2%) and the additive , dominant , and recessive models , showed a significant association between rs179247 and a decreased susceptibility to GD for the total group (2 = 76.5%) and the additive , dominant , and recessive models  showed a significant association between rs179247 and a decreased susceptibility to GD (12192 = 0%) and the additive , dominant , and recessive models , showed a significant association between rs12101255 and an increased susceptibility to GD for the total population group (2 = 0%) and the additive , dominant , and recessive models  showed a significant association between rs12101255 and an increased susceptibility to GD . In his study, Brand did not state the detailed ages and genders, which would be a source of heterogeneity2 = 0%, P = 0.927) and for Caucasians , and the pooled OR was still statistically significant for both the total group  and for Caucasians . We found some heterogeneity in all of the genetic models for GO rs179247, though these results were not significant (I2 ≤ 45.1%). There were three different population groups for GO rs179247, including Caucasian, Chinese and Brazilians that could be the source of heterogeneity.For the total population groups, there was significant heterogeneity for the Allelic model of GD rs179247 . We also assessed the studies by ethnicity, and the results showed that heterogeneity was still significant for Caucasians , the percentage of female cases (p = 0.493) and the sample size of cases (p = 0.953).We performed a sensitivity analysis by excluding each study once in all of the genetic models . The pooGraves’ disease is a common autoimmune disease with genetic susceptibilityFor the total group and the Caucasian group, our results showed that the carriers of rs179247 GG had a 34% less risk of developing GD than the controls. Rs12101255 TT carriers had 26% and 23% more risk of developing GD than the controls in the total group and the Caucasian group, respectively. Allele A of rs179247 and allele T of rs12101255 were more frequent in GD patients. Allele G of rs179247 and allele C of rs12101255 were more common in healthy people.The location of rs179247 and rs12101255 within the first intron, which is close to the promoter region and the start codon, may influence post-translational processes or gene expression. As the only confirmed disease-specific gene, two mechanisms have been proposed to explain the association of TSHR intron 1 SNPs with the risk of developing GDThe first mechanism is the alternative splicing process and generation of soluble TSHR . The splicing of exons code for the extracellular domain of the protein is influenced by intronic TSHR polymorphisms. The intronic bases do not exist in the mature mRNA, but they are important in regulating function. The intronic SNPs can influence mRNA splicing and change protein functions. The main transcriptions of TSHR encode 3 different TSHR: the full-length TSHR (flTSHR), ST4 and ST5. ST4 and ST5 have the same 8 exons as flTSHR and an additional ninth exon. Two regions of intron 8 encode the ninth exon so that the introns were retained. In the Brand study that we included in the meta-analysis, the author evaluated flTSHR, ST4 and ST5 expression in 12 thyroid tissue samplesThe second mechanism is the modulation of TSHR expression in the thymus . The risk alleles result in a lower TSHR expression in the thymus, which means that less TSHR self-reactive T cells will be deleted. More TSHR-Abs can be produced by the germinal centres of the thyroid-draining lymph nodes. Colobran identified this hypothesis in his studyTSHR rs179247 and rs2268458 polymorphisms had no association between GO (GD with GO) and GD (GD without GO) for any of the genetic models. In the Jurecka study, which we included in the meta-analysis, the rs179247 TSHR polymorphism had a genetic association with GO for young patients only (age of onset ≤30 years)TSHR is the auto antigen for both GD and GO. The pathogenesis of GD/GO is complex and multifactorial with many environmental and genetic factors. Environmental factors, such as smoking and old age, are considered risk factors for GO. However, the genetic associations of TSHR between GD and GO are still unclear. Compared with GD (GD without GO) patients, no SNPs were associated with GO (GD with GO) in our meta-analysis even though these SNPs were positively associated with GD. Because 25–50% of GD patients have a clinical manifestation of GO, more studies are needed to identify aetiological factors besides the gene.As far as we know, our study is the first meta-analysis to evaluate the genetic association of TSHR polymorphisms with GO. We searched all the available data with no language limits. Studies of high quality were selected with inclusion and exclusion criteria. We also performed sub-group analysis, meta-regression analysis and sensitivity analysis to test the reliability of the statistical results. Hardly any publication unfairness was identified by either Begg’s funnel plot or Egger’s regression test. As such, the data analysis was stable and reliable. The central and tissue-specific role of TSHR gene was further confirmed with GD/GO in our meta-analysis.Some limitations of the genetic studies in GD/GO were revealed. First, Caucasian and Asian people were the primary study population groups. Other ethnic information was limited. More ethnic data are needed for further studies. Second, the number of TSHR genetic studies in GO was small. Therefore, more studies are needed to identify the genetic associations of TSHR with GO. Third, genetic studies in GD/GO should collect more evidence of the environmental, hormonal and antigenic factors and the interactions of these factors with genetic factors. These data would strengthen our understanding of the multi-factorial pathogenesis of GD/GO.In summary, we identified TSHR rs179247 and rs12101255 as genetically associated with GD. TSHR rs12101255 and rs2268458 polymorphisms had no genetic association between GD and GO. Future studies with bigger sample sizes and different ethnic population groups are needed to confirm the genetic association of TSHR gene polymorphisms with GD/GO. In addition, TSHR genetic studies should investigate gene-environment as well as gene-gene interactions. This approach would lead to a more precise understanding of the genetic association of TSHR gene polymorphisms with GD/GO.We searched Pub Med and EMBASE for case-control studies published up to December 2015 that studied the genetic association of TSHR gene polymorphisms with GD/GO. The search strategy was based on a combination of “(thyroid stimulating hormone receptor gene OR TSHR gene) AND  AND ”.We also manually scanned the references in articles and reviews to include all potentially relevant articles with no language limitationsThe inclusion criteria were as follows: (a) studied the association of TSHR gene polymorphisms with the GD/GO; (b) was a case-control study; (c) had sufficient published data for estimating an odds ratio (OR) with a 95% confidence interval (CI); and (d) diagnosis of GD was based on standard clinical criteria, including increased serum levels of free thyroxine (FT4) and triiodothyronine (T3), decreased TSH, positive TRAb values, presence of a diffuse goitre as detected by ultrasonography, and increased thyroidal uptake of pertechnetate. The eye examinations included proptosis measurement; measurement of eyelid width; assessment of muscular function; assessment of corneal status; examination of the fundus oculi; and measurements of visual acuity. The exclusion criteria were as follows: animal studies, case reports, reviews, abstracts, conference proceedings, editorials or studies with incomplete data.Two investigators (H.B.X. and M.X.W.) screened and reviewed each article and completed the data extraction independently. Disagreements were resolved by group discussion with a third investigator (X.Y.Z.). To study the genetic associations of the TSHR gene polymorphisms with GD/GO, we selected the most strongly associated SNPs to analyse . Data were extracted by two investigators (H.B.X. and M.X.W.), and the data included the first author, publication year, location, population ethnicity, definition and sample of the cases and controls, gender and mean age of cases, and the DNA extraction and genotype results. The Hardy-Weinberg equilibrium (HWE) was used to test the genetic equilibrium2 value. Statistically, a larger amount was considered when P < 0.05. If the P value was ≥0.05 or the I2 value <50%, a fixed-effect model was used, and then a random-effect model was adoptedhttp://www.ohri.ca/programs/clinical epidemiology/oxford.asp)Each SNP was used for meta-analysis if there were 2 or more studies. Different genetic models were used to assess the genetic association, including the allelic (B vs. A), dominant (AB + BB vs. AA), recessive (BB vs. AA + AB) and codominant  modelsAll calculations were performed using Stata 12.1 .How to cite this article: Xiong, H. et al. Genetic associations of the thyroid stimulating hormone receptor gene with Graves diseases and Graves ophthalmopathy: A meta-analysis. Sci. Rep.6, 30356; doi: 10.1038/srep30356 (2016).


The media have a key role in communicating advances in medicine to the general public, yet the accuracy of medical journalism is an under-researched area. This project adapted an established monitoring instrument to analyse all identified news reports (n = 312) on a single medical research paper: a meta-analysis published in the British Journal of Cancer which showed a modest link between processed meat consumption and pancreatic cancer. Our most significant finding was that three sources  appeared to account for the content of over 85% of the news stories which covered the meta analysis, with many of them being verbatim or moderately edited copies and most not citing their source. The quality of these 3 primary sources varied from excellent  to weak , and this variance was reflected in the accuracy of stories derived from them. Some of the methods used in the original meta-analysis, and a proposed mechanistic explanation for the findings, were challenged in a subsequent commentary also published in the British Journal of Cancer, but this discourse was poorly reflected in the media coverage of the story. Accurate media coverage is essential to maintain public trust in science and medicine. The public mainly learn about advances in science and medicine through the mainstream media  , althougDiet and dietary advice are particular areas where research has found problems in the media coverage of medicine and science. A study conducted in the UK found that less than a third of all dietary advice given in UK newspapers (across one week) was supported by levels of evidence that were considered satisfactory . Diet isWe conducted a case study of one scientific publication; a meta-analysis examining whether there is a relationship between consumption of red/processed meats and the development of pancreatic cancer . PancreaThe meta-analysis studied here, by Larsson and Wolk from the Karolinska Institute in Sweden, analysed 11 studies and found, from 7 of them, a small but significant relationship between the self-reported consumption of processed meat and the risk of developing pancreatic cancer. Specifically, the self-reported daily consumption of 50g of processed meat was associated with a 19% increase in risk. In the discussion, the authors speculated on possible mechanistic explanations for the association they observed, including a putative relationship between the consumption of nitrites/nitrates and the risk of developing cancer . The LarNews stories were identified by internet searches using Google, Google News, and the Nexis database. Searches were carried out between November 2012 and September 2013. Search terms were derived using basic information from the meta-analysis (see terms below). If one news story then identified further relevant stories (e.g. as sources or links) then these were also included. Search terms were entered into Google and the search results analysed page-by-page until the researcher was satisfied that no further stories could be found (by searching through the results until no relevant stories had been returned for at least 5 screen pages).'british journal cancer processed meat cancer' “an extra 50g of processed meat every day increases a person’s risk by 19%”‘Larsson Wolk pancreatic cancer’“11 trials and 6,643 patients with pancreatic cancer”‘Karolinska Institute pancreatic cancer’‘Lightsey pancreatic cancer’‘Eating one sausage daily could raise the risk of pancreatic cancer by fifth, found a research’“meat ‘linked to pancreas cancer’”“The sample sizes ranged from 17,633 to 1,102,308, and the number of pancreatic cancer cases that occurred in the studies varied from 57 to 3,751”Phrases used to search Google/Nexis:The last two phrases were used specifically to search for stories which reproduced a story first seen on the NHS choices site, once it became apparent that many had done so.Focus was specifically the Larsson and Wolk meta-analysis, interpretation and analysis of findingsEnglish languageAccessible onlineWritten (rather than audio or video)Not a forum/social media (e.g. Facebook/Twitter) posting or other ‘chat site’News stories had to meet the following criteria to be included in the analysis:Stories from ‘conventional’ news outlets were included, as well as the websites of trade journals, health sites, blogs and any other site which might reasonably be considered to be ‘reporting’ on Larsson and Wolk. The full list of 312 stories analysed is given in Criteria were developed by two authors (JT and PN) using the established Health News Review/Media Doctor criteria as the starting point ,5,24. ThWe did not consider it possible or useful to try and objectively make a distinction between the individual criteria regarding their relative importance for public understanding see also . For thiStories which satisfied a criterion were scored ‘1’, those which did not were scored ‘0’, thus a higher score was reflective of a ‘better’ story. Based upon the original analysis methods developed by the ‘Media Doctor’ system, each news story was independently analysed against the criteria by a minimum of two raters from the study team, using the guidance provided above, with any scoring conflicts resolved either through discussion between the raters or by the use of an additional rater ,5,25.Criteria for analysis . Every story was assessed against every criterion. For each criterion, the analysis proceeded to ask does the story;1adequately quantify the increased risk of pancreatic cancer? Pancreatic cancer is rare; the lifetime risk is approximately 1 in 77 for men and 1 in 79 for women, so a 19% increase in risk is actually very little. To satisfy this criterion, a news story had to report—or make some attempt to explain—the absolute risk of getting pancreatic cancer.2grasp the quality of evidence? The Health News Review [s Review  use this3not represent general 'disease mongering'? Potential examples could include—sensationalist headline, omission of key facts to increase the perception of risk, generalise risk to all cancers. A failure to satisfy other criteria could also result in a perception of disease mongering (e.g. failure to report absolute risk) but those would not result in a zero score for this criterion (i.e. they should not be double counted).4use independent sources? i.e. not from the author(s), or the journal where the meta-analysis was published , or Cancer Research UK .5not represent a potential, unacknowledged conflict of interest? e.g. is from an anti/pro-meat eating/industry publication.6discuss/acknowledge that the Larsson and Wolk study did not control for variables such as body-mass index and a history of diabetes? (From Lightsey [Lightsey )7discuss/acknowledge difficulties with using recall and epidemiological data to make accurate predictions about risk? (From Lightsey [Lightsey )8discuss the limitations of linking the Larsson and Wolk results to a hypothesis about nitrites? Larsson and Wolk presented a link between nitrites and pancreatic cancer as a ‘plausible’ mechanistic explanation for their findings. This was challenged by Lightsey [Lightsey . To sati9draw in other findings and present a bigger picture? Did the article put the news story into context of other findings on a similar topic—e.g. other studies looking at pancreatic cancer, or the effects of processed/red meats on other cancers. This criterion is derived from the Health News Review criterion suggesting stories  should ‘establish the true novelty of the treatment’ and ‘compare the new treatment with existing alternatives’.10acontain a significant amount of original journalism and not represent a cut-and paste from another source? This related directly to the journalism, i.e. was there sufficient additional material to say that the news story is more than a simple copy from another source. Getting a ‘1’ for this did not necessarily mean that the rater thought journalist had read the paper itself. Some stories appeared to be copied directly from another source and then edited down, perhaps with the odd extra line/statement. These would also score ‘0’.10bIf ‘0’ to 10a, is the source of the material cited?11adequately credit Larsson and Wolk? To satisfy this criterion, the story had to give sufficient information for readers to understand who had conducted the study and where it was published. (e.g. link to/describe source of paper).In addition, records were also made of the length of the news story (words) and the date published (where available).The length of individual stories was determined by copying the text of the story into Microsoft Word and using the ‘word count’ feature. In a small number of cases, the ability to copy text from a website was restricted and so words were counted directly off the screen.312 news stories were identified, including reports from major national and local newspapers and other news sources around the world, as well as health sites and independent blogs  which did not meet criterion 10  revealed that all but one were derived largely or wholly from 1 of 3 sources. These sources were journal press release, BBC (British Broadcasting Corporation) News and NHS Choices. The host institute for the Larsson and Wolk study also issued a press release, but this appeared to be derived from the journal press release. A comparison of the performance of these stories with the performance of their source story is shown in Table 4 shows a comparison between the press release/BBC-derived stories and those which appeared to be original. Although we did not consider it valid to conduct a detailed statistical comparison of the performance of these groups , articles appearing there are written by the company ‘Bazian’ and then contracted out to primary care centres, thus it seems likely that the NHS Choices/Bazian site is the source for these stories, including those appearing in the aforementioned local newspapers. All 137 of these stories were 1667 words long and met 10 of the 11 criteria.As described above, 137 stories were identical verbatim copies of each other, appearing on the NHS Choices  website, and 91 local newspapers in the UK and Ireland, mostly owned by the Johnson Media group and the Newsquest Media group. The story also appeared in the news sections of 43 primary care centres in the UK. According to the NHS Choices Website , have found that approximately one third of them are derived from journal press releases . The curA finding of inaccurate media coverage of a science story relating to dietary advice is unfortunately not new, although the area remains under researched . PreviouThe causes of these concerns are varied, and there is a growing recognition that the responsibility for ensuring accurate media coverage lies with all those involved, not just journalists . The preOne common disconnect between science and the media coverage of it relates to the issue of ‘balance’. Many broadcasters are committed to media coverage that is ‘balanced’, a notable example being the UK BBC which is funded by the UK taxpayer and has strict rules on impartiality . UnfortuHowever, the present study has also demonstrated the power of a good primary news source. The article appearing on the ‘NHS Choices’ website and in 136 other locations, was of high quality and independently addressed, in part, some of the points raised by David Lightsey. Additionally, 14.5% of stories ‘went beyond a secondary source’ altogether and, in doing so, approximately one third of them also addressed 2 of the 3 main points raised by Lightsey, again despite being published before the Lightsey commentary. We did not find any news stories which cited the Lightsey commentary and at the time of writing (January 2015) the Lightsey commentary has not been cited in any scientific publication . In addition to the recommendations discussed above regarding additional guidelines, transparency and media training, these findings suggest that, in the short term, one simple solution to the ‘weak link’ in the chain connecting science and the media (i.e. the press release) would be for media organisations to make greater use of independent and well researched sources such as NHS Choices, themselves identified through research such as that presented in the current study.There are some limitations to the current study. Despite extensive searching, we cannot claim to have identified all the news stories whose focus was the Larsson and Wolk meta-analysis. Furthermore we have no data which would have allowed us to determine how many people have read the individual stories, although the list of stories analysed includes major news organisations around the world see . We madeWe based our analysis on published criteria for the assessment of news stories regarding science and medicine ,5,24, whscientific content was informed only by a secondary source was higher than the 266 identified . For example, the BBC news story contained only scientific information that is also present in the press release, with additional journalism and comments from an independent source. Similarly, some of the 46 news stories identified as ‘primary’ still use the BBC as a source for some of their content but added enough original journalistic content to satisfy criterion 10; thus the reach of the BBC story was greater than the 40 stories which are largely based upon it. This last issue is part of a larger limitation which affects all studies of this type – the need for subjective decision making when assessing news stories against criteria such as those used here. The impact of this limitation was minimised through various means, for example by basing our analyses on existing, established criteria [Finally, it is difficult for a study of this nature to accurately trace the scientific content of the original study. For the 46 stories  that ‘were not based largely on secondary sources’, it is not possible for us to conclude that the authors of those stories have actually read the original scientific manuscript. This raises the possibility that the number of stories whose criteria  and ensucriteria ,25 and icriteria ), althoucriteria ) while ocriteria ,39).In summary, we identified 312 distinct news stories, all reporting on a single meta-analysis describing a potentially increased risk for pancreatic cancer associated with processed meat consumption. Over 85% of the news stories were based on one of three primary sources, which themselves varied in quality, amplifying the net effect of any errors or weaknesses in those sources. This emphasises the need for rigour and accuracy in scientific and medical press releases, which are the indirect source of public education about advances in science and medicine.S1 FileEvery story analysed in the study is listed, along with a link to the webpage (where possible).(XLSX)Click here for additional data file.


VEGF) gene polymorphisms and stroke. However, their findings remain controversial. The objective of this study was to evaluate the relationships between VEGF gene polymorphisms and stroke by a meta‐analysis.Numerous studies have investigated the relationships between vascular endothelial growth factor  with 95% confidence interval (CI) were calculated.The PubMed, Embase, China National Knowledge Infrastructure database, Wanfang Chinese database, and ORs and 95% CIs of +936 T were 1.44, 1.09–1.90, P = 0.01  and 1.19, 0.85–1.65, P = 0.31, and the overall ORs and 95% CIs of −1154 A were 0.98, 0.87–1.10, P = 0.75 and 0.95, 0.82–1.11, P = 0.53. No publication bias was found in this meta‐analysis.Ten studies were included, including a total of 2331 cases and 1814 controls for +936C>T, 3040 cases and 2649 controls for −1154G>A. Under the dominant and recessive models, respectively, the overall The meta‐analysis showed that +936C>T may be a risk factor for stroke, especially in Asians, while −1154G>A was not associated with stroke. Stroke is one of the most common causes of death and the most common cause of long‐term disability in the world, which endangers human health and increased the burden on society  is a prime angiogenic factor and a regulator of endothelial cell proliferation. The VEGF gene is located on chromosome 6p21, which is comprised of a 14 kb coding region with 7 introns and 8 exons  Medline in PubMed searching engine; (2) Embase database; (3) Chinese National Knowledge Infrastructure (CNKI) database; (4) Wanfang Chinese database; and (5) VIP database. The latest date for searching literatures was November 1st, 2015. The key words for literature searching were: [“vascular endothelial growth factors” or “vasculotropin” or “VEGF”] and [“stroke” or “cerebral infarction” or “cerebrovascular disorders”] and [“single nucleotide polymorphism” or “SNP” or “polymorphism” or “mutation” or “genetics” or “variant”]. Publication language was restricted to English and Chinese, and the subjects were not limited in our search. We also performed a manual search of the reference lists of retrieved articles to find additional potential studies.The inclusion criteria for the gene association studies in this meta‐analysis were as follows: (1) independently published studies explored the association between VEGF gene polymorphisms and stroke; (2) with genotype or allelic distributions provided; (3) with data in any of the two polymorphisms, and sufficient data available to calculate an odd ratio (OR) with its 95% confidential interval (CI); (4) if the authors published two or more studies using the same subjects, the most recent publication or the publication with the largest sample size was included. No limitations were placed on race, ethnicity, or geography area.Relevant data were systematically extracted from the included studies by two authors using a standardized form, and reached a consensus on all items. The researchers collected the following data: the first author's name, publication year, countries and ethnicities of participants, sample size, and genotyping method.To determine the methodological quality of the included studies, we used the Newcastle–Ottawa scale (NOS)   of the distribution of controls was checked by Pearson's Figure P = 0.01) and 1.19, 0.85–1.65 (P = 0.31) compared with C in the dominant and recessive models, respectively . The overall ORs and 95% CIs of −1154 A were 0.98, 0.87–1.10 (P = 0.75) and 0.95, 0.82–1.11 (P = 0.53) compared with G in the dominant and recessive models, respectively , except a relatively high heterogeneity in dominant model of +936C>T (I2 = 73%). Given this we performed subgroup analysis and meta‐regression to investigate the sources of heterogeneity. Age, gender, ethnicity, sample size, genotyping method, publication language were tested, the results confirmed the sources to be none of them   Fig.   and −1154G>A (rs1570360), which are all established to influence VEGF expression. The relationship between two polymorphisms and stroke remains obscure due to inconsistent results. To our knowledge, up to now, no such meta‐analysis has been performed to investigate the relationship between VEGF gene polymorphisms and the risk of stroke. Whether the two polymorphisms can affect the risk of stroke is concerned in this study particularly.P = 0.01), which was more significant in Asians . However, the data indicated no relationship between −1154G>A and stroke, in both models.In the present meta‐analysis, we investigated the relationship between two VEGF gene polymorphisms and stroke susceptibility. Data extracted from published studies were combined to estimate genetic associations between the most commonly investigated polymorphisms of VEGF, +936C>T and −1154G>A, and stroke. The results of the meta‐analysis indicated that one of the two polymorphisms, +936C>T, may be associated with the increased risk of stroke. Individuals with +936 T showed increased risk of stroke in dominant model , IL‐1 and matrix metalloproteinases (MMP) (Li and Qin The current meta‐analysis has limitations that should be acknowledged. First, the number of studies enrolled in this meta‐analysis was relatively small. Some null study findings may be not published. Second, all included articles were published in English or Chinese. Therefore, studies issued in other languages might be missed. Third, there was heterogeneity in the dominant model of +936C>T, and we failed to find the source of the heterogeneity by subgroup analysis and meta‐regression. Even though we used the random effects model to reduce its affect, it still makes the results uncertain. Fourth, as the number of included studies was relatively small and some of the original articles didn't provide more details of the subtypes, we failed to perform stroke subtypes analysis. In addition, we analyzed five genetic models, allele model, dominant model, recessive model, additive model, and codominant model, there were significant findings only in the dominant model.In conclusion, our results suggested that the +936C>T polymorphisms may be associated with the risk of stroke, especially in Asians, but not −1154G>A polymorphisms. And more studies in this field are still needed to make the association more conclusive.The authors report no conflicts of interest. The authors alone are responsible for the content and writing of the paper.


In the developed countries, the incidence of esophageal adenocarcinoma (EAC) is increasing over recent decades. The purpose of this meta-analysis was to arrive at quantitative conclusions about the contribution of alcohol intakes and the progression of Barrett's esophagus.A comprehensive, systematic bibliographic search of medical literature published up to Oct 2013 was conducted to identify relevant studies. A meta-analysis was conducted for alcohol consumption on the Barrett's esophagus progression.A total of 882 cases in 6,867 individuals from 14 observational studies were indemnified in this meta-analysis. The result of this current meta-analysis, including 10 case-control and 4 cohort studies, indicated that alcohol consumption was not associated with the neoplastic progression in Barrett's esophagus . When stratified by the study designs, no significant association was detected in either high vs low group or ever vs never group.Alcohol drinking is not associated with risk of neoplastic progression in Barrett's esophagus. Further well designed studies are needed in this area. The incidence rate of Barrett's esophagus in the patients with gastroesophageal reflux disease (GERD) was 10% to 15% Helicobacter pylori status, presence and duration of GERD symptoms, and proton pump inhibitor (PPI) use and duration. However, the risk factors of the neoplastic progression in the Barrett's esophagus remained unclear. Alcohol, as a risk factor of several kinds of cancers Factors that may be associated with an increased risk of Barrett's esophagus or EAC were examined, including: age, gender, ethnicity, smoking, alcohol use, Barrett segment length, hiatal hernia presence and size, Meta-analysis is a useful statistical tool to pool the relevant studies together and gain a more powerful conclusion. The meta-analysis was also used in the search for potential causes of EAC. For instance, based on a combination of 4 cohorts and 31 case-control studies, Salehi M et al found that low levels of red and processed meat consumption and higher levels of fish intake might reduce esophageal cancer risk As a meta-analysis of observational studies, this current meta-analysis was conducted according to the PRISMA guidelines and MOOSE guidelines Studies satisfying the following criteria were included in the observational meta-analysis: (1) the association between alcohol consumption and risk of Barrett's esophagus progression; (2) a cohort or case-control study design was obtained; (3) the value of relative risk (RR), odds ratio (OR) with 95% confidence intervals (CI) or enough data to calculate them were reported.Two reviewers (ZL and HX) independently extracted the data and any discrepancy was checked again and resolved through discussion. The following data was extracted from each article: name of the first author, publishing year, study design, study site, age and gender of participants, type and amount of cases, adjustments of the confounding factors, and the OR/RR value with 95% CI.Quality assessment that was performed by two reviewers back to back and any disagreement was discussed with the third reviewer. Considering the observational study design of the included studies, the Newcastle-Ottawa Scale (NOS) was obtained to assess the methodological quality of the included studies 2 statistic. When P for the heterogeneity <0.1 and 2I>50%, the interstudy heterogeneity would be considered statistically significant. Both OR and RR were reported in the included studies and the OR was obtained to approximate RR in this meta-analysis. When both crude and adjusted OR/RR values were offered in the article, only the adjusted ones were adopted for the meta-analysis. A random-effects model was obtained to estimate the pooled effects, thus it would provide a more conservative conclusion. The effects of alcohol drinking on the incidence of Barrett's esophagus progression were measured with the OR with 95% CI. The sensitivity analyses were conducted to detect the robustness of the conclusions by two independent methods. Firstly, we conducted a sensitivity analysis to investigate the influence of a single study on the overall risk estimate by omitting one study in each turn. Secondly, we excluded the studies with a relative lower methodological quality and assess the effect of alcohol consumption and risk of progression of Barrett's esophagus.Homogeneity of RR across studies was assessed by using the Cochrane Q statistic and IWe constructed a funnel plot with logRR and their SEs of logRR by visual inspection to assess the potential publication bias. Besides, potential publication bias was also assessed by both Begg's rank correlation test A total of 433 articles (198 from Pubmed and 235 from EMBASE) were identified from the electronic database searching. Besides, 131 more records were identified through consulting the reference lists of the relevant reviews and articles. After excluding 382 articles with unrelated topics, a total of 182 records were detailed evaluated. Among the 182 articles, 31 full-texts were assessed for eligibility after removing 151 articles . Subsequently 2 articles were duplicated reports from previous data and 15 ones in which the data not in usable format were excluding from the inclusion and in final, a total of 14 studies were included for the quantitative synthesis A total of 882 cases and 6,867 individuals in 14 observational studies were indemnified in this current meta-analysis. All the detailed characteristics of each included study were presented in The NOS were obtained to assess the selection, comparability and exposure of the case-control study, while the selection, comparability and outcome of the cohort study. All the scores of each part in the evaluation of all the studies were displayed in 2I, 41.3%; P = 0.063). We tried to explore the source by excluding the included studies one by one and re-count the heterogeneity. When one study 2I, 9.2%; P = 0.769). The result of this current meta-analysis didn't change when that study was excluded .The heterogeneity was significant when all the 14 studies were pooled in the meta-analysis .A one-way sensitivity analysis was conducted and there was little change in the quantitative summary measure of RR or 95% CI. Although one studies P = 0.381; Egger's test, P = 0.645).Evidence of publication bias for studies in this current meta-analysis wasn't noted in symmetrical funnel plot on visual inspection . Both BeA total of 882 EAC cases in 6,867 individuals from 14 observational studies were identified in this meta-analysis. The result of this meta-analysis, including 10 case-control and 4 cohort studies, indicated that alcohol consumption was not associated with the neoplastic progression in Barrett's esophagus. Meanwhile, this result was demonstrated in the most subgroup analyses by study design, study sites, end points. However, the studies conducted in the Americas showed that alcohol drinking increased the neoplastic risk. When stratified by the study designs, no significant association was detected in either high vs low group or ever vs never group. The heterogeneity was not significant when all the 14 studies were pooled in the meta-analysis. Publication bias was not detected in the meta-analysis. The results of the sensitivity analysis suggest that the conclusions of this study were quite robust.EAC is now a more and more serious problem in the entire world In this meta-analysis, we found that alcohol consumption is not associated with progression of Barrett's esophagus. This result supports the conclusions of several previous studies. In 2003–2004, a prospective, multicenter cohort study including 713 patients with Barrett's esophagus was conducted. After 4 years of follow-up, duration of Barrett's esophagus of ≥10 years, length of Barrett's esophagus, esophagitis, and LGD were significant predictors of progression to HGD or EAC; however, alcohol intake was not associated with the progression of Barrett's esophagus . In a hospital-based case-control study in which 91 cases with EAC and 244 controls with histologically confirmed Barrett's esophagus (>2 cm) with no dysplasia or low-grade dysplasia were included, current alcohol use was not associated the incidence of EAC  However, the studies conducted in the Americas showed that alcohol drinking was a risk factor of the progression of Barrett's esophagus. The geographical differences, the diet diversity and ethnic and genetic disparity are the possible reasons. Alcohol consumption, which was classified as beer, liquor and wine intake, might demonstrate different effect in the development of Barrett's esophagus. The difference of the drinking habits in each region might cause the results. Besides, the relatively small number of studies included in the subgroup analyses (5 in the Americas and 1 in the Africa) might lead to the instability of the conclusions. When stratified by the end points, no significant association was detected in either HGD or EAC or just EAC. It suggests that alcohol might be not associated in any stage of the Barrett's esophagus progression.To our best knowledge, this is the first meta-analysis involving the relationship between alcohol consumption and risk of development of Barrett's esophagus. There are some strengths of this work. In the literature search, a large number of subjects were evaluated for the detection of the effect of the neoplastic risk in Barrett's esophagus associated with alcohol drinking. Besides, through two independent methods, the publication bias wasn't significant. The results of the sensitivity analysis suggest that the conclusions of this study were quite robust. These above results demonstrated that the conclusions of this meta-analysis were quite persuasive.recent studies revealed that cardiac-type mucosa, columnar lined esophagus without goblet cells, also have the posibility of cancer development. With the development of understanding of Barrett's esophagus, the older papers might provide results based on inaccurate definitions and thus the bias could not be ignored. Advanced well-designed studies based on both endoscopic results and histocytology are wanted in the future.Despite these advantages mentioned above, some limitations of the current meta-analysis should be acknowledged. First, the definitions of the case groups were not uniformly defined. Both the patients without higher alcohol intake or ever alcohol intake were obtained as the controls in the included studies. This would produce potential misclassification bias. While the subgroup analyses showed that no different results were detected and thus this might produce no serious problem. Secondly, even the categories of alcohol consumption are reported in some studies; however, the data were varied and broad. Thus, we were unable to determine the dose-response association between alcohol consumption and progression of Barrett's esophagus. Third, it has been argued that because meta-analyses of observational studies may produce very precise, but spurious, results, a statistical combination of these data should not be the prominent component. However, considering that as an etiology exploring nature of this study, the pooled results of the observational studies would also provide certain improvement of the knowledge of the neoplastic development. The end, a combination of prognostic factors has been suggested to be required to define subgroups of patients at an increased risk of progression. However, the data of the included studies reported no sufficient data for an advanced research, which demonstrated the potential breakthrough. The forth, the understanding of the Barrett's esophagus has developed in the recent years. The old conception says that although several different kinds of columnar epithelium can be found in the lower esophagus, it is only specialized intestinal metaplasia (SIM), distinguished by the presence of goblet cells, which increased the cancer risk. However, nowadays In conclusion, on the basis of epidemiologic evidence, we found that alcohol consumption was not a neoplastic risk in Barrett's esophagus. To get a more definitive conclusion, further pooled analyses with more complete raw data or prospective cohort studies with larger sample size, well controlled confounding factors and longer duration of follow-up are needed in this area.Figure S1The funnel plot of all the included studies. Evidence of publication bias for studies in this current meta-analysis wasn't noted in symmetrical funnel plot on visual inspection.(TIF)Click here for additional data file.Checklist S1PRISMA checklist.(DOC)Click here for additional data file.


The aim of this review and meta-analysis was to critically determine the possible effects of different cooling applications, compared to non-cooling, passive post-exercise strategies, on recovery characteristics after various, exhaustive exercise protocols up to 96 hours (hrs). A total of n = 36 articles were processed in this study. To establish the research question, the PICO-model, according to the PRISMA guidelines was used. The Cochrane’s risk of bias tool, which was used for the quality assessment, demonstrated a high risk of performance bias and detection bias. Meta-analyses of subjective characteristics, such as delayed-onset muscle soreness (DOMS) and ratings of perceived exertion (RPE) and objective characteristics like blood plasma markers and blood plasma cytokines, were performed. Pooled data from 27 articles revealed, that cooling and especially cold water immersions affected the symptoms of DOMS significantly, compared to the control conditions after 24 hrs recovery, with a standardized mean difference (Hedges’ g) of -0.75 with a 95% confidence interval (CI) of -1.20 to -0.30. This effect remained significant after 48 hrs  and 96 hrs . A significant difference in lowering the symptoms of RPE could only be observed after 24 hrs of recovery, favouring cooling compared to the control conditions . There was no evidence, that cooling affects any objective recovery variable in a significant way during a 96 hrs recovery period. For decades, different cryotherapies–such as cold-water immersions (CWI) and ice packs–have been used for post-exercise recovery in a variety of sports to cope with fatigue and/or delayed-onset muscle soreness (DOMS) . ExercisIn sports medicine, cryotherapy, as a post-exercise intervention, has been investigated by means of subjective ratings of DOMS and general ratings of perceived exertion (RPE), objective attenuations of blood plasma markers such as creatine-kinase (CK) and lactate-levels, or blood plasma cytokines encompassing Interleukines (IL) and C-reactive protein (CRP) , 2, 4–37Population: healthy, non-injured, female and male study participants, participants could be of any training status; Intervention: external applications, only post-exercise cryotherapy accepted; Comparator: passive control interventions without cooling, passive sham / placebo applications; Outcomes: DOMS and RPE, levels of venous or capillary blood plasma markers (CK-levels and blood/muscle lactate-levels respectively lactate dehydrogenase [LDH]-levels) and/or blood plasma cytokines , measured 24 hrs, 48 hrs, 72 hrs and/or 96 hrs after exhaustive exercise. DOMS and RPE were defined as subjective recovery characteristics in this study. The subjective perception of DOMS was rated on various scaling systems, such as visual analogue scales (VAS), BORG scales or Likert-scales. RPE was rated by various scaling systems, such as BORG scales or VAS. CK-levels, lactate-levels, CRP and IL-6 were defined as objective recovery characteristics in this study. The rationale for inclusion of these outcome variables was comparability. Numerous researchers who tried to identify a subjective and objective recovery profile after exhaustive exercises previously used these variables in their studies.The research question was defined by the PICO-model in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement : PopulatA systematic search was accomplished electronically between October 2013 and August 2014 in the MEDLINE (PubMed), SportDiscus and PEDro databases according to the PRISMA statement . The folInclusion criteria for the articles were: (1) the study design was randomized or quasi randomized into an intervention group and a non-cooling, passive control group; (2) the cold therapy had to be an external application form without combinations, focused on the mentioned outcome variables after the exercise bouts; (3) the cold interventions had to be compared to at least one control condition; (4) the outcome variables had to be measured immediately (0–24 hrs), and / or 24 hrs, and / or 48 hrs, and / or 72 hrs and/or 96 hrs post-exercise; (5) the data of the outcome variables had to be processed in the articles; (6) the human volunteers had to be healthy without any physical infirmity; (7) the human volunteers could be of any athletic training status; (8) there were no sex-defined inclusion or exclusion criteria; (9) at least one of the subjective or objective recovery characteristics had to be evaluated in the articles; (10) the cold therapy had to be applied within one hour after the end of the exercise protocol; (11) all exhaustive exercise types or muscle damaging protocols were accepted; (12) English and German language restrictions. Articles, which did not meet these criteria were excluded from this study. 2,were calculated to assess across studies heterogeneity and its degree respectively [General characteristics of the studies and results of the individual studies were extracted independently by two researchers . In case of a lack of information, the statistics were extrapolated from the given figures and graphs. In case of disagreement, consensus was sought by involving a third researcher (JPB). Meta-analyses with a random-effects model (specified a priori), allowing for possible between–studies heterogeneity, were used to examine the overall effect size of cooling applications on the subjective and objective recovery characteristics. The inversed variance method was used to calculate the weighting factors for the included, individual study results. The different individual study effect sizes were standardized using Cohen’s d, by dividing the effect sizes by the pooled standard deviations, as proposed by Borenstein et al. (2009) . The effectively . If betwectively , 8. To aectively . The ComIn this article the subjective variables under investigation were DOMS and RPE. The objective outcomes of interest for the meta-analyses were blood plasma markers  and blood plasma cytokines . Meta-analyses were conducted using means ± standard deviations (SD) or standard errors (SE).The systematic error of the 36 articles was assessed using the Cochrane’s risk-of-bias tool. . Two resThe risk of bias analysis demonstrated a high risk of bias for the blinding procedures. Blinding of the participants was only described in two studies , 9. PersA total of n = 36 articles met the inclusion criteria. These trials were processed in the present article as sources of primary research data, while the systematic review articles and meta-analyses were used in the introduction and the discussion part. Characteristics of the included studies are summarized in The exercise protocols encompassed soccer-games or tournaments , 7. Six In the present article, the most common cold therapy application (28 studies) was CWI of the legs –37, 50. WBC with sequential entering -110°C, -60°C and -10°C cold chambers was used in two studies . One aut-1) snack, a carbohydrate bar (PowerbarTM) and 600 ml of fluid in form of a sports drink (GatoradeTM). We decided not to exclude this study because of the stretching protocol. In all studies the control and experimental groups received interventions of the same average duration.The predominant passive control condition, used in 30 studies, was sitting, standing or resting supine in a room with comfortable temperatures, ranging from 15°C to 24°C –37, 50. We obtained the data from 27 articles which revealed the effectiveness of cooling on DOMS , 34, 35.2: 78.4%). The classic fail and safe model analysis showed, that 317 negative studies would be needed before alpha increases above the 5% level for DOMS, measured 24 hrs after the cold application, indicating a low risk of publication bias. Cold therapy continued to attenuate the symptoms of DOMS significantly up to 96 hrs : 19, p < 0.001; I2: 73.3%); : 11, p = 0.013; I2: 54.3%); : 7, p = 0.238; I2: 24.0%). Up to 72 hrs, a moderate to very high heterogeneity was observed [2: 82.0%); : 14, p < 0.001; I2: 77.9%); : 8, p = 0.018; I2: 56.7%); : 6, p = 0.233; I2: 25.7%). These results indicate, that the findings in favour of cooling are dependent on the cooling modality. Subgroup analysis according to sex showed significant different effects, favouring cooling for male participants compared to female participants, up to 48 hrs : 18, p < 0.001; I2: 75.2%); : 16, p < 0.001; I2: 66.9%). To analyse the effect of the different cooling temperatures on the attenuation of the symptoms of DOMS, a meta-regression analysis between these parameters was performed. The meta-regression for the cooling temperature across the recovery time showed no significant correlation between cooling temperature and relieving symptoms of DOMS . Visual analysis of the regression showed two extreme low temperatures used in the study from Hausswirth et al. (2011) and Costello et al. (2012) [Only two of these studies used a pain pressure algometer to measure DOMS , 26. Six2: 81.7%). The meta-regression according to cooling temperature could not explain this high heterogeneity, nor show a significant correlation between cooling temperature and Hedges’ g . Two cooling modalities  were used to assess the effect of cooling on RPE. The subgroup analysis according to cooling modality showed no significant differences between the cooling modalities, measured after 24 hrs of recovery : 1, p = 0.477). No significant differences between cooling and the control condition could be observed after 48 hrs of recovery. : 2, p = 0.044; I2: 67.9%).In this composed article, data of n = 6 studies were pooled up to 48 hrs for the analysis of the subjective RPE , 24, 32.The effect of post-exercise cooling on various blood plasma markers was analysed in 26 studies –37, 50. 2: 58.3%). The classic fail and safe model analysis suggested, that 6 studies would be needed to bring alpha above the 5% level, indicating a potential risk of publication bias. Subgroup analysis according to time (< 24 hrs vs. 24 hrs) and sex could not explain the heterogeneity. However, the cooling temperature and also the diagnostic procedure to obtain blood lactate or muscle lactate results could significantly explain parts of the heterogeneity . Tucker et al. (2012) were the only authors who analysed the lactate-levels by measuring local muscle lactate [2: 0%). A meta-regression according to cooling temperature showed no significant differences between cooling and alleviated lactate-levels during the 24 hrs recovery time .Eight studies analysed blood lactate-levels , 37, 50,et al. 20 were the2: 63.8%; 48 hrs: Hedges’ g: -0.19, 95% CI: -0.46 to 0.08; Cochran’s Q: 18.4, df (Q): 13, p = 0.143; I2: 29.4%; 72 hrs: Hedges’ g: -0.10, 95% CI: -0.46 to 0.27; Cochran’s Q: 10.3, df (Q): 7, p = 0.175; I2: 31.7%). The subgroup analyses could not significantly explain the high heterogeneity, which was only found after 24 hrs of recovery. A meta-regression according to cooling temperature was performed at 24 hrs. The results showed no significant slope between cooling temperature and Hedges’ g . After excluding the extreme low temperatures, presented in the study from Hausswirth et al. (2011), the results of the meta-regression still did not change significantly  [The possible effects of post-exercise cooling on the CK-levels was investigated in 19 studies up to 72 hrs , 33, 35.Ten studies used plasma cytokines (IL-6 and CRP)–amongst other–as a recovery variable , 28, 33.2: 0%; 48 hrs: Hedges’ g: -0.35, 95% CI: -0.85 to 0.15; Cochran’s Q: 0.9, df (Q): 2, p = 0.646; I2: 0%; 72 hrs: Hedges’ g: -0.43, 95% CI: -0.93 to 0.07; Cochran’s Q: 0.54, df (Q): 2, p = 0.764; I2: 0%). Similar results could be observed for the CRP values. No significant results favouring cooling compared to the control conditions could be observed during the 24 hrs recovery period : 6, p < 0.001; I2: 76.3%). The meta-regression for cooling temperature showed no significant slope after 24 hrs of recovery . However, there were significant differences after 48 hrs of recovery, favouring cooling compared to the control condition : 3, p = 0.075; I2: 56.5%). A subgroup analysis according to cooling modality or cooling temperature was not possible because a meaningful grouping was not applicable.There were no significant main effects, favouring cooling compared to the control conditions during the 72 hrs recovery period, according to the IL-6 levels : 3, p = 0.473; IThe majority of the studies had a high or unclear risk of bias, which made the interpretation of the results uncertain. The main bias was produced by means of the blinding procedure. The possibilities for blinding cooling applications stay limited. Another problem was the description of the random sequence generation and the procedure of allocation concealment. The power of the selected articles was mainly small according to the individual study population. Bleakly et al. (2012) examined the effects of CWI for preventing and treating muscle soreness after exercise and reported similar limitations according to the individual study population . Another2: 67.88%). A possible explanation could be that the well-trained volunteers can provide more power and more DOMS and RPE than the normal trained volunteers. These findings can be useful for physiotherapists, coaches and physicians who work with athletes. Especially athletes, who perform exhaustive exercises or sports and possess only short recovery periods between training sessions or competitions, could benefit from these findings.One main finding of this study was, that cold therapy (CWI) significantly alleviated the symptoms of DOMS 24 hrs, 48 hrs and 96 hrs after the cooling application. This significant effect was not present at 72 hrs after the cooling application. By performing subgroup analysis, sensitivity analyses and meta-regressions, we could not detect the reason for this non-significant result. However, Tseng et al. 2013) showed a trend, favouring the sham application compared to cooling application . These a showed a2: 56.52%) according to Cohen et al. (1992) and Higgins et al. (2002), which could not be explained by meta-regressions [2: 58.30%). All meta-regressions according to cooling temperature indicated, that the cooling temperature did not correlate to Hedges’ g. Especially very low temperatures (-110°) which were used in cold chambers, seem to negatively influence the effect of cooling. One needs to consider that very low cooling temperatures were only used for a short period or without direct skin contact. Gregson et al. (2011) recently showed that CWI induced a reduction of femoral artery blood flow and muscle temperature after a 10 min (8°C to 22°C) immersion [Cooling did not affect objective recovery variables such as lactate-levels, CK-levels or IL6-levels. However, significant differences could be observed favouring cooling at CRP 48 hrs. Because of the high risk of bias according to the blinding procedures and the unclear selection bias, it is not certain that these results represent the real effects of cooling at this point. Furthermore, only four studies with a total of n = 65 volunteers represent this significant outcome. No inflammatory item was significantly lower at any time in this meta-analysis except the CRP value measured at 48 hrs, which made this result questionable. Although the effect size was moderate (Hedges’ g: -0.72), a high heterogeneity could be observed  and RPE (up to 24 hrs) compared to passive control interventions. CWI achieved the best effect with respect to the other cooling applications. To sum up the results of the individual studies: the mean temperature of the studies, showing a significant result favouring cooling compared to the passive recovery intervention, was 10°C (range: 5°C to 13°C). The reported and suggested cooling time for alleviating the subjective symptoms is 13 min (range: 10 min to 24 min). Cooling did not significantly affect the objective recovery outcomes compared to passive control interventions. Despite the detailed analysis of the individual study results, it must be viewed with caution, because the risk of bias was high according to the blinding procedures and remained unclear for the selection bias. The between-studies heterogeneity could be partly explained. Other factors (various pre-cooling exercises and various passive non-cooling recovery strategies) might influence the different recovery characteristics.S1 PRISMA Checklist(PDF)Click here for additional data file.


Interleukin-4(IL-4) is a critical inflammatory cytokine and has been involved in pathogenesis of cancer. To date, several studies have investigated the association between IL-4 intron 3 variable number of tandem repeats (VNTR) polymorphism and cancer risk in humans; however, the results remain controversial. We performed this meta-analysis to find a more conclusive association between this polymorphism and cancer risk.Eight eligible case–control studies were identified through searching electronic databases, including 1583 cases and 1638 controls. Odds ratio (OR) and corresponding 95% confidence interval (CI) were used to estimate the strength of the association.The results of overall analyses indicated that the variant RP2 allele was associated with a decreased cancer risk compared with the RP1 allele . In subgroup analyses stratified by ethnicity, there was evidence in the Asian population for an association between this polymorphism and cancer risk .IL-4 intron 3 VNTR polymorphism could influence the risk of human cancer. Due to the limitations of this meta-analysis, further well-designed and functional researches should be performed to validate our results. Cancer is currently a major health burden in the world which results from complicated interactions between genetic and environmental factors ,2. EpideInterleukin-4(IL-4), produced by activated Th2 type CD4+ T cells, represents a key differentiation cytokine that induces development of Th2 subset of lymphocytes, which is responsible for surveillance and clearance of tumor cells by activation of granulocytes and eosinophils, as well as inhibition of angiogenesis ,7. MoreoThe gene encoding IL-4 is located on chromosome 5q31.1 . A rapidTo date, several case–control studies have been carried out to explore the linkage between IL-4 intron 3 VNTR polymorphism and the risk of several human cancers. However, results from different articles remain controversial. We performed this meta-analysis based on the published studies to make a more conclusive evaluation of the association between this polymorphism and cancer risk.Relevant articles indexed in PubMed and Cochrane Library  were independently searched by two authors, using the key words as follows: (“interleukin-4” or “interleukin 4” or “IL-4” or “IL 4”) and  and (“polymorphism” or “polymorphisms” or “SNP” or “variant” or “variation”). The retrieved results were filtered to English language papers. An additional manual search was performed among the references of relevant articles and related articles in PubMed.All the retrieved articles were reviewed by two authors independently to select studies for inclusion. As for studies with overlapping data reported by the same investigators, the articles with the most complete data were eligible. Studies included in this meta-analysis should meet the following predetermined criteria: 1) case–control design, 2) evaluation of the IL-4 intron 3 VNTR polymorphism and cancer risk, 3) effective estimation of odd ratio (OR) with 95% confidence interval (CI), or enough data to allow calculation of these two statistics. The major criteria for exclusion were: 1) not relevant to IL-4 polymorphism and cancer risk, 2) not with a case–control design , 3) study on inherited cancers, 4) duplicate data.Two authors checked each included article independently to extract the necessary data. The following information were collected: first author’s last name, publication year, ethnicity of the study population, country where the study conduct, sample size, source of the control groups , the distribution of variant genotypes of both cases and controls.Pooled analysis was performed to estimate the strength of the association between IL-4 polymorphism and cancer risk using the odds ratio (OR) with 95% confidence interval (CI). The pooled ORs were calculated by a co-dominant model , a dominant model (RP2/RP2 + RP1/RP2 vs. RP1/RP1), a recessive model (RP2/RP2 vs. RP1/RP1 + RP1/RP2), and an allelic model (RP2 vs. RP1). The values of the pooled ORs were tested by Z-test . Hardy-W2 statistic [2 < 50%) or presence (I2 > 50%) of heterogeneity. Sensitivity analyses were conducted by removing one study each time to assess the stability of the results. The potential publication bias of the included studies was evaluated by Begg’s funnel plots graphically and Egger’s linear regression test [Heterogeneity among the included studies was evaluated using chi-square based on Q-test and Itatistic . Pooled tatistic ) or randtatistic ) effectiion test .All the tests were two sides and it is deemed to be statistically significant when P < 0.05. All the statistical analyses were carried out with Stata/SE software version 12.0 .Our search in electronic databases identified 444 records. Four hundred and one records were excluded after reviewing titles and abstracts and we gained 43 useful studies. After full-text reviewing, we excluded 35 studies and listed the reasons for their exclusion in Figure There were 8 studies –33 suitaThe main results of this meta-analysis were presented in Table In addition, we omitted 1 study in which the control group was not in agreement with HWE  and repe2 = 0.00% for heterogeneity). Moreover, the pooled ORs in all genetic models were not altered qualitatively when any single study was removed, which indicated the results of the present meta-analysis were relatively stable and credible . Sensitivity analyses were carried out to explore the origin of heterogeneity. Two studies ,33 were The publication bias of eligible literatures was assessed by Begg’s funnel plots graphically and Egger’s test statistically. The shapes of funnel plots in all models did not show any evidence of an obvious asymmetry Figure . MeanwhiCancer is a heterogeneous disease resulting from genetics, environmental factors, and their interactions, the clinical outcome of which is difficult to predict –36. InflTo our knowledge, Li et al. performed a meta-analysis about the association of IL-4 intron 3 VNTR polymorphism and cancer risk . Their rThe present meta-analysis, including 1583 cases and 1638 controls from 8 case–control studies, comprehensively assessed the association between IL-4 intron 3 VNTR polymorphism and risk of human cancers. We found an association between IL-4 intron 3 VNTR polymorphism and cancer risk. Our results indicated the RP1 allele might be a risk factor for malignancies. The results were robust because the pooled ORs did not alter in sensitivity analyses. Stratified analyses suggested that the association was mainly in Asian populations and in bladder cancer. Since there was significant heterogeneity in several genetic models, two studies were founded out by sensitivity analyses. We re-checked these 2 studies and found that 1 study  was the Some limitations of our meta-analysis should be pointed out. First, the pooled outcomes were calculated based on unadjusted estimates, which limited a more precise evaluation on adjusted estimates by several important factors like age, sex, lifestyle and etc. Only one study reported that IL-4 intron 3 VNTR polymorphism was associated with oral and pharyngeal carcinoma risk, which interacted with alcohol drinking . Thus, lDespite these limitations, advantages in our meta-analysis should be also acknowledged. First, the statistical power was definitely increased by pooling a substantial number of cases and controls. Second, all the eligible studies met the inclusion and exclusion criteria strictly and completely. Third, no publication bias was detected through Begg’s funnel plots and Egger’s test, which indicated that the pooled results should be unbiased.In conclusion, the results of this meta-analysis were statistically credible. The relationship between IL-4 intron 3 VNTR polymorphism and cancer risk was assessed and this polymorphism was associated with cancer risk, mainly in Asians. To draw a more precise conclusion, further studies should be carried out with more detailed individual information, concerning the effects of other polymorphisms and haplotypes, enrolling larger sample size of cases and well-matched controls, especially in Caucasians and Africans, to validate the role of IL-4 polymorphism in carcinogenesis.


SLC2A9 and ABCG2 and CHD risk in a Chinese population.Multiple studies investigated the associations between serum uric acid and coronary heart disease (CHD) risk. However, further investigations still remain to be carried out to determine whether there exists a causal relationship between them. We aim to explore the associations between genetic variants in uric acid related loci of SLC2A9 and rs4148152 in ABCG2) and CHD risk was performed by logistic regression model. Adjusted odds ratios (ORs) with 95% confidence intervals (CIs) were calculated. Compared with subjects with A allele of rs4148152, those with G allele had a decreased CHD risk and the association remained significant in a multivariate model. However, it altered to null when BMI was added into the model. No significant association was observed between rs11722228 and CHD risk. The distribution of CHD risk factors was not significantly different among different genotypes of both SNPs. Among subjects who did not consume alcohol, the G allele of rs4148152 showed a moderate protective effect. However, no significant interactions were observed between SNP by CHD risk factors on CHD risk.A case–control study including 1,146 CHD cases and 1,146 controls was conducted. Association analysis between two uric acid related variants  contains supplementary material, which is available to authorized users. Coronary heart disease (CHD) is one of the leading causes of morbidity and mortality throughout the world . The WorUric acid, as the end product of purine metabolism, is a major cause of gout ,9. StudiSLC2A9  [ABCG2  [SLC2A9 and ABCG2 and the risk of CHD among Europeans [Recent genome-wide association studies (GWASs) identified multiple genetic loci associated with serum uric acid concentrations ,18,26,27 4p16.1) ,26,28,292, 4q22) ,30,31 po2, 4q22) . Severaluropeans ,33,34 anSLC2A9) and rs4148152 (intron in ABCG2), which were in the previously reported uric acid related loci and confirmed in our GWAS [In the present study, a case–control design  was adopted and two SNPs rs11722228  in Wuhan city  between May 2004 and October 2006. Quantitative coronary angiography was performed by experienced cardiologists who had no knowledge of the patients’ clinical information. After exclusion of those who had acute renal and liver diseases or with incomplete information, 2,292 individuals of 1,146 cases with 1,146 age- and sex- frequency matched controls were included in our study. All the subjects were unrelated Chinese Han individuals and lived in Hubei province, the central China. The baseline characteristics of cases and controls were shown in Table General health examination was performed including standing height and body weight. Height was measured to the nearest 0.01 cm with subjects standing without shoes. Weight was measured using a digital scale with subjects wearing light clothing and recorded to the nearest 0.1 kg. Body mass index (BMI) was calculated as body weight in kilograms divided by standing height in meters squared ,37. ThosThe diagnostic criteria for CHD cases included one of the followings: (1) presence of a stenosis > 50% in at least one of the major segments of coronary arteries  based on coronary angiography, which can be seen in more details in previous studies ,40,41;  anticoagulation tubes for plasma and DNA. The blood specimens were frozen in −80°C until assayed. Genomic DNA was isolated with a Puregene kit .SLC2A9) and rs4148152 (ABCG2), which were associated with uric acid levels in our recent GWAS [P > 0.05) except for a slight deviation of rs4148152 in controls (P = 0.02) (data not shown).Two SNPs rs11722228  unless otherwise specified. We conducted logistic regression analysis to calculate adjusted ORs and their 95% CIs for CHD risk by different genotypes of these SNPs in the multivariate models. Multivariate model 1 included age, sex, smoking, drinking, and family history of CHD. Multivariate model 2 included the same set of variables in model 1 plus BMI. Based on the model 2, model 3 further included total cholesterol, triglyceride, and the history of hypertension and diabetes. Homozygous genotypes CC and AA were used as reference genotypes for rs11722228 and rs4148152, respectively. The interactions between the independent SNPs and the covariates such as age, sex, BMI, smoking, and drinking were tested by introducing the SNP × environmental factor terms into the multivariate logistical regression model. Simultaneously, general linear model was performed and P for trend was calculated to observe the distribution of several traditional CHD risk factors among different genotypes of both SNPs. A two-side P value of < 0.05 was considered statistically significant. All statistical analyses were performed by the statistical analysis software package SPSS 12.0.Categorical variables were presented in percentages and compared by Chi-square analysis. Continuous variables were expressed in mean ± SD and compared by student’s P < 0.05). However, the levels of TC and LDL cholesterol were significantly lower in cases than those in controls, which might be due to the intake of cholesterol-lowering medications in CHD cases. The percentage of past history of diabetes and hypertension, and family history of CHD in cases were dramatically higher in contrast to that in control subjects .In the present case–control study, CHD controls n = 1,146) were frequency matched for age and sex to cases . Baseline characteristics of study individuals are shown in Table ,146 wereP = 0.006) and the association stayed significant in a multivariate model  adjustment for age, sex, smoking, drinking and family history of CHD (model 1). However, the association altered to null when BMI was introduced into the model (model 2). Subjects carried the G allele of rs4148152 had a decreased risk of CHD (AG + GG) . The results remained significant in model 1  but changed to borderline significant in model 2 when BMI was introduced into the model (P = 0.062). Further adjustment for the remaining traditional risk factors including total cholesterol, triglyceride, and past history of hypertension and diabetes got the similar null results (P = 0.306).As shown in Table 2), smoking status (yes/no), alcohol consumption status (yes/no). Table P = 0.038), however, no significant interactions were observed between alcohol consumption and SNP of rs4148152 on CHD risk (P for interaction = 0.47). Similarly, no interactions were observed for the two SNPs and other covariates on CHD risk.We further investigated the associations between these two SNPs and CHD risk stratified by several traditional CHD risk factors such as sex , BMI  [ABCG2) [It still remains to be determined whether serum uric acid is an independent risk factor of CHD risk. As Mendelian randomization indicates, genetic variants could serve as an instrument to explore the causal associations between the biomarkers and the risk of diseases ,42. In tSLC2A9) ,26,28,29 [ABCG2) ,30,31, w [ABCG2) , to explSLC2A9 and ABCG2 loci to be positively associated with serum uric acid levels and gout [SLC2A9, also known as GLUT9 (glucose transporter type 9), is a glucose transporter and plays a significant role in maintaining glucose homeostasis. SLC2A9 is a causative gene for renal hyperuricemia and plays a significant role in urate reabsorption on renal proximal tubular cells [ABCG2 is one of adenosine triphosphate (ATP) binding cassette family and expressed in kidney proximal tubule cellular membrane [Previous GWASs have identified and gout ,22,43. Sar cells . ABCG2 imembrane ,44. It tUnfortunately, we did not find significant associations between the two variants and CHD risk. In addition, considering that the CHD traditional risk factors might modify these associations, we conducted stratification and interaction analysis but no significant interactions were found between these covariates and the two variants on CHD risk. The results indicated that the associations of the uric acid related variants with CHD risk were not modified by these CHD traditional risk factors and there might be no causal association between serum uric acid levels and CHD risk.P < 0.05. However, the present study only had 32% power to detect variants with MAF = 0.3 and OR = 1.1 at P < 0.05 significant level. Further studies with larger sample size were needed to validate our results. Thirdly, the present study was conducted in Chinese Han population, further studies conducted in other populations were necessary.Several issues contributing to this null result should be noted. Firstly, only two independent uric acid related variants SNPs of rs11722228 and rs4148152 were selected to perform this association study and they explained only 1.03% and 1.09% of the total variation of serum uric acid levels, respectively . SelectiSLC2A9 with the risk of CHD in a Chinese population. Subjects carried the G allele of rs4148152 in ABCG2 locus had decreased CHD risk, however, this association altered to borderline significant when BMI and other traditional risk factors were introduced into the multivariable model. No significant interactions between the two SNPs and CHD related risk factors were observed. Studies with larger sample size in other populations and genotyping more variants related to uric acid levels were warranted in future studies.In summary, this study did not find significant association of uric acid related SNP rs11722228 in