Ribeiroia. We used geographic information system (GIS) land-use/land-cover data to validate field observations of land use in the proximity of study wetlands. In a multiple logistic regression model that included land use as well as developmental stage, genus, and water-quality measures, proximity to agricultural land use was associated with an increased risk of limb malformation . The overall discriminant power of the statistical model was high (C = 0.79). These findings from one of the largest systematic surveys to date provide support for the role of chemical toxicants in the development of amphibian limb malformation and demonstrate the value of an epidemiologic approach to this problem.The cause of limb deformities in wild amphibian populations remains unclear, even though the apparent increase in prevalence of this condition may have implications for human health. Few studies have simultaneously assessed the effect of multiple exposures on the risk of limb deformities. In a cross-sectional survey of 5,264 hylid and ranid metamorphs in 42 Vermont wetlands, we assessed independent risk factors for nontraumatic limb malformation. The rate of nontraumatic limb malformation varied by location from 0 to 10.2%. Analysis of a subsample did not demonstrate any evidence of infection with the parasite  In 1995, amphibians with severely malformed limbs were discovered in Minnesota . The nexThe exact etiology of amphibian limb abnormalities  remains unclear. The major suspects have been ultraviolet B (UV-B) radiation, trauma, parasitic trematode infestation, and xenobiotic pesticides or chemicals . AlthougIn laboratory settings, some evidence has supported each of the etiologic hypotheses. Controlled exposure to UV radiation has produced amphibian dysmorphogenesis , 2002 anAnother primary etiologic hypothesis of limb abnormality is trauma related to predation. Although predation can account for some forms of abnormality in the wild, the hypothesis, more generally, has not been supported as the primary cause of increased amphibian abnormality rates in recent years . AlthougRibeiroia ondatrae. Infestation with Ribeiroia is associated with limb abnormalities in some amphibian species  and four species of ranids  were collected from 42 wetlands in the Lake Champlain Basin of Vermont. We selected wetlands in a representative fashion along an urbanization gradient ranging from relatively undisturbed forest habitat , to rural communities, to neighborhoods in Burlington. To collect specimens at each site, standardized collection methodology consisted of pipe samplers  and dip nets (46 cm × 23 cm with 1-mm mesh). An attempt was made to collect up to 300 specimens from each site (a maximum of 100 during each of three visits). After capture, specimens were first anesthetized in a solution of tricaine methanesulfonate and then placed in 70% ethanol, according to protocols for the study that were approved by the Yale University Animal Care and Use Committee. Each specimen was later examined in the laboratory in order to determine genus and species, Gosner stage, and presence of malformation. A subsample of specimens was examined by dissection for the presence of parasite metacercariae, including Ribeiroia.Between 24 May and 28 August 2002, amphibian specimens consisting of two species of hylids (p < 0.0001 for both). Both observed agriculture and lawns were negatively correlated with GIS land-use estimates for forest cover (p < 0.0001 for both).For each of the 42 wetlands, on-site field observations were used to determine whether agriculture or lawns were located proximate to the wetland. The presence of agriculture and lawns were then scored as yes/no responses for each site. In addition, we independently estimated land use/land cover within 200 m of each amphibian sample site, which was calculated using geographic information system (GIS) data. The latitude and longitude coordinates of each site were used with land-use/land-cover GIS layers to map the land use within 200 m of each site. The Raster land-cover data set from Vermont GIS U.S.  was brouSeveral quantitative measurements were made at each site, including pH, conductivity, dissolved oxygen, temperature, total nitrogen, and total phosphorus for the water samples. If more than one measurement took place at an individual wetland, the measurements were averaged together for that site.Ribeiroia infection.As an additional investigation into the possible role of parasitic infection, a representative sample of snails from the study wetlands was investigated for evidence of Each individual specimen was examined by a trained technician and classified according to a case definition adapted from a well-established amphibian limb abnormality classification scheme . To be cFor the analysis of limb malformations, we included only those specimens with Gosner stage of ≥26 . At thisOf the original 5,983 specimens, this led to the exclusion of 684 with stage < 26, and 27 with missing stage, leaving 5,272. An additional 8 cases were excluded because of unknown length, leaving 5,264 in the final study sample. For our case definition, we excluded trauma deformities  because of the differing etiologies of trauma and nontrauma abnormalities.All analyses were performed using SAS software .Simple frequencies were calculated for categorical variables, including Gosner stage, genus, water-quality measures, presence of pollution sources, and presence of limb malformation. Means and SDs were calculated for continuous variables.r > 0.40) were excluded from the multivariable analyses. We found that a number of variables had significant collinearity, including nitrogen and phosphorus (r = 0.67) and pH and dissolved oxygen (r = 0.79). Therefore, neither phosphorus nor pH was included in the final multivariate model.To determine bivariate associations between exposures and rates of malformation, we performed simple logistic regression. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated for each individual factor. An OR calculates the odds of a risk factor being present in the affected cases compared with the odds that it is present in the non-affected individuals. A multivariate logistic regression model was developed to determine the independent predictors of malformation. To determine which variables to include in the multivariate model, we examined collinearity between independent variables using correlation coefficients. Variables with correlations above levels at which multicollinearity would be considered likely (p > 0.05). ORs, 95% CIs, and p-values for association between individual risk factors and limb malformation were determined for variables remaining in the final model. The overall discriminant power of the model was assessed using a C-statistic.The multivariate logistic regression analysis used a backward selection process to eliminate nonsignificant variables from the model (criteria for elimination from the model set at n = 40) revealed no evidence of Ribeiroia infection. Similarly, representative samples of host snails from the sample wetlands did not demonstrate evidence of Ribeiroia infection .Examination of a subsample of individual specimens (p < 0.0001). In other words, the risk of malformation increased 18% for each increase in Gosner stage. Although malformation rates were slightly higher in ranids than in hylids, the effect of genus was not significant in the multivariate model. Furthermore, analyzing risk factors for malformation in ranid species alone or hylid species alone produced results similar to those of our combined model (data not shown). In the multivariate model, proximity to agriculture remained a highly significant predictor of malformation risk . None of the other independent variables tested showed a significant association in the multivariate model. The discrimination and fit of the logistic regression model were good, with a C-statistic = 0.79 and a p-value for the Hosmer-Lemeshow goodness-of-fit test of 0.10.In the multivariate logistic regression model, Gosner stage remained highly significantly associated with malformation . Evaluation of multiple stressors will require continued development of effective study methodologies and a movement away from the single-agent hypothesis testing of the past . To our Rana pipiens tadpoles to extracted compounds from semipermeable membrane devices in freshwater ponds could cause malformation in the presence of UV radiation  was lower than figures reported for amphibian limb abnormalities in some other surveys . First, The study relied on a limited number of measurements of water quality, as well as visual observations of land use as an indicator of potential pollution sources. Therefore, the observed associations between runoff sources, measured water quality, and malformation rate must be viewed as preliminary. Determining exposure at the level of the group rather than the individual risks the “ecologic fallacy” of assigning potential causation to one factor when in fact another, unmeasured factor is actually responsible. Although the rate of malformation in ponds near agriculture was increased, it could be that this difference was due to other factors that were not assessed. Further studies involving direct measurements of toxicants in water, and case–control comparisons of affected and unaffected individuals need to be carried out in order to identify candidate toxicants or chemical mixtures that could be investigated in the laboratory setting.Water-quality measures are dependent on multiple factors and vary over a season and year to year depending on runoff sources, rainfall, temperature, and other factors. By using a limited number of water-quality measures taken during a single year, it is possible that some exposure misclassification occurred, with water conditions during sensitive periods of amphibian development varying from the measurements at the time of sampling. However, such misclassification, in epidemiologic studies of cause and effect, tends to be nondifferential and bias the results toward observing no effect . TherefoThe human health relevance of these findings remains to be determined. If a chemical toxicant or mixture could be identified that is capable of causing limb malformations in wild amphibians, several additional lines of investigation could shed light on whether a risk to humans exists. First, it would be useful to know what human health outcome would be analogous to amphibian limb deformities; this could be explored if gene sequence homology between humans and amphibians was present for target genes affected by the toxicants. Second, it would be important to establish a dose–response relationship and determine whether environmental exposures to such a toxicant would be capable of causing human health effects. Finally, epidemiologic studies including both amphibian and human populations could determine whether rates of malformation in amphibians show a correlation with a suspected human health outcome. Although to date none of these connections have been established, they will be worthy of further exploration if confirmatory studies support the role of toxicant chemicals in the etiology of amphibian developmental abnormalities.


The base deficit is a useful tool for quantifying total acid–base derangement, but cannot differentiate between various aetiologies. The Stewart–Fencl equations for strong ions and albumin have recently been abbreviated; we hypothesised that the abbreviated equations could be applied to the base deficit, thus partitioning this parameter into three components .The two abbreviated equations were applied retrospectively to blood gas and chemistry results in 374 samples from a cohort of 60 children with meningococcal septic shock . Partitioning required the simultaneous measurement of plasma sodium, chloride, albumin and blood gas analysis.r2 = 0.83, y = 1.99 – 0.87x, standard error of the estimate = 2.29 meq/L). Hypoalbuminaemia was a common finding; partitioning revealed that this produced a relatively consistent alkalinising effect on the base deficit (effect +2.9 ± 2.2 meq/L (mean ± SD)). The chloride effect was variable, producing both acidification and alkalinisation in approximately equal proportions ; furthermore the magnitude of this effect was substantial in some patients (SD ± 5.0 meq/L).After partitioning for the effect of chloride and albumin, the residual base deficit was closely associated with unmeasured anions derived from the full Stewart–Fencl equations (It is now possible to partition the base deficit at the bedside with enough accuracy to permit clinical use. This provides valuable information on the aetiology of acid–base disturbance when applied to a cohort of children with meningococcal sepsis. Metabolic acidosis is a common biochemical finding in critically ill patients . The proRecent insights into acid–base physiology (the Stewart–Fencl approach) have provided a method of quantifying each component of the acid–base status ,7,11. HoThe study was approved by the Institutional Ethics Committee, which waived the need for informed consent.We examined data retrospectively from 68 consecutive patients with meningococcal sepsis admitted to the paediatric intensive care unit from January 2001 to June 2003. Cases were identified from the departmental database. Patients with meningococcal meningitis without septic shock were excluded. Septic shock was defined as the need for more than 40 ml/kg of fluid resuscitation within 4 hours of presentation to hospital or the requirement for inotropic medication . All bloAfter exclusion of those without septic shock, full data were available for 374 blood samples from 60 patients (giving a median of six samples per patient). Patient demographics were as follows: median (interquartile) age 2 years (0.8 to 9.5), weight 13 kg (10 to 19), Paediatric Index of Mortality version 2 (PIM2)-derived mortality risk 11.0% (6 to 16), crude mortality 10.0% (PIM2-predicted death rate 13.8%). In addition, 88% of patients required mechanical ventilation, 82% received inotropes, and the amount of fluid administered in the first 24 hours after admission was 158 ± 65 ml/kg (mean ± SD).tot) was calculated according to the algorithm for 'standard base deficit' in the blood gas analyser, which necessitated the concurrent measurement of haemoglobin. Plasma albumin was measured by binding of bromocresol green dye, and whole blood lactate by the enzymatic method . The precision values for the above analysers were as follows: pH 0.009 to 0.005 , pCO2 2.74 to 2.78%, ion-specific electrodes all less than 2%, albumin 6.2% and 3.0% , and lactate 2.0%.Arterial blood gases and chemistry were measured with the Instrumentation Laboratory 1640 blood gas analyser  and Synchron LX20  The total base deficit ; strong ions, of which chloride concentration (relative to sodium) is the most important (BDcl); and net unmeasured anions from tissue acids (BDUMA). Blood lactate can be considered as either a strong ion (if measured) or an unmeasured anion (if unmeasured). For the purposes of this study we designated lactate as an unmeasured anion.BDThese three factors can exert an acidifying  or an alkalinising  effect on the total base deficit, according to their net charge ,7,11.If formulae quantifying the effect of albumin and chloride on the base deficit are applied and subtracted from the total base deficit, the remainder should equal the contribution from net unmeasured anions , so thattot - BDalb - BDCl = BDUMABDUMA) should therefore approximate unmeasured anions calculated from the Stewart–Fencl equations. After considering the precision of the blood gas and chemistry analysers, the expected loss of accuracy due to the abbreviated nature of the base deficit equations, and the normal value for the Stewart–Fencl strong ion gap (see below), we set an a priori limit of 3 meq/L for the standard error of the estimate (SEE).If this method is accurate, the residual ] × 0.25BDCl = [Na+] - [Cl-] - 32BDA full explanation of the Stewart–Fencl methodology is reported elsewhere ,7,11,15;Unmeasured anions (strong ion gap) = strong ion difference (measured) - strong ion difference (effective)+ + K+ + Ca2+ + Mg2+] - [Cl- ] Strong ion difference (measured) = [NaCO2/(10-pH)) + 10 × [Alb] (g/L) × (0.123 × pH - 0.631) + [PO42-] (mmol/L) × (0.309 × pH - 0.469).Strong ion difference (effective) = (12.2 × pUMA) and Stewart–Fencl methods (strong ion gap) was assessed by linear regression with the use of the ordinary least-squares method (Microsoft Excel).Data are reported as means and SD. Agreement between unmeasured anions calculated from the base deficit (BDtot -7.4). The unmeasured anion-related base deficit was greater than the total base deficit; this was predominantly due to the alkalinising effect of hypoalbuminaemia , SEE 2.29 meq/L.Not surprisingly, BDn = 60). Measurements were chosen by means of the random number generator in Excel using a uniform distribution, whereby the sample with the largest assigned random number from each patient's group of samples was chosen. This produced remarkably similar results: r2 = 0.854, Stewart–Fencl-derived UMA = 2.39 - (0.871 × BDUMA), SEE 2.17 meq/L. We therefore conclude that the above approach is valid.Finally, it is noted that the regression analysis used multiple samples taken from each patient; thus each measurement cannot be considered truly independent in a statistical sense . In this setting, multiple measurements taken on an atypical patient could potentially bias the regression analysis. To investigate this we reanalysed the data in two ways. First, a standardised residual plot was inspected, which did not reveal obvious deviation from normality among the residuals, nor any extreme outliers. Second, we repeated the regression analysis, using one measurement only per patient (r2 = 0.83). However, to assess whether this model is accurate enough for clinical use, we must consider three other aspects of the regression analysis, namely the SEE (2.29 meq/L), the slope (-0.87) and the intercept (1.99 meq/L).These findings demonstrate that the base deficit can be partitioned at the bedside by the application of two simple formulae, requiring the measurement of plasma sodium, chloride and albumin concurrently with the arterial blood gas. This was validated by comparing the unmeasured anion portion of the base deficit with that calculated from the Stewart–Fencl equations, yielding a high coefficient of determination  did not reveal an unusual pattern; furthermore, the residuals seemed normally distributed with consistent variance. Thus we can say that about 95% of the time the true unmeasured anions will lie within ± 4.5 meq/L of that estimated by the partitioned base deficit (1.96 × standard error).The sources of this error are threefold, including both the abbreviated albumin and chloride equations and the fact that phosphate, the other major weak acid, is not accounted for. Albumin charge is a function of pH ,11,16,17However, this effect is not large; Story has estimated a typical error to be less than ± 1 meq/L , which iThe slope of the regression equation (-0.87) can be interpreted as meaning that a decrease in base deficit of -10 meql/L will represent an increase of 8.7 meq/L in unmeasured anions; in essence this is close enough to be considered as an inverse equimolar relationship. It is also notable that the intercept, which occurs when the base deficit is equal to zero, yields an estimated unmeasured anion value of 2 meq/L, which is consistent with the normal value for this parameter .In summary, we feel that the properties described above permit bedside partitioning of the base deficit, provided that the user is aware of the limitations and sources of error. Partitioning provides the clinician with valuable information about the aetiology of an acidosis, which can have implications for treatment and prognosis. Examples are outlined below.Underestimation of tissue acidosis: critically ill patients have a high incidence of hypoalbuminaemia that produces an alkalinising effect, masking the true degree of 'tissue acidosis' ,17,18. IRecognition of iatrogenic causes of an acidosis: the use of albumin solutions for resuscitation is common in paediatrics, and may become more widespread in adult practice since the publication of recent safety data . AlbuminPrognosis: the prognostic value of an acidosis related to unmeasured anions is uncertain, with studies producing conflicting results -30; thisA potential criticism of the partitioning approach is that it may offer the same information as the anion gap. This is partly true, provided that the anion gap is corrected for albumin ,18. HoweUMA into lactate and non-lactate components. This is important, because the two components are not tightly correlated [This study did not attempt to address the role of lactate, but merely sought to validate a method for partitioning the base deficit. The prognostic and therapeutic value of lactate measurement is well established ,32,33; trrelated ,9.In summary, we have validated two simple equations that permit partitioning of the base deficit into three components , providing for a more detailed bedside analysis of acid–base disturbances. We have found this to be useful in everyday clinical practice.• 	It is possible, by application of two simple equations, to partition the base deficit into three components: chloride, albumin and unmeasured anions.• 	This requires simultaneous measurement of an arterial blood gas, and venous plasma sodium, chloride and albumin.r2 = 0.83) in a cohort of patients with meningococcal sepsis.• 	Agreement between unmeasured anions calculated from the partitioned base deficit and from the full Stewart–Fencl equations produces good agreement ).• 	The effect of chloride on the base deficit was more variable, producing significant acidifying and alkalinising effects in almost equal measure (effect -0.5 ± 5.0 meq/L (mean ± SD)).alb = base deficit due to albumin; BDCl = base deficit due to chloride; BDtot = total base deficit; BDUMA = base deficit due to unmeasured anions; PIM2 = Paediatric Index of Mortality version 2; SEE = standard error of the estimate.BDalb = base deficit due to albumin; BDCl = base deficit due to chloride; BDtot = total base deficit; BDUMA = base deficit due to unmeasured anions; PIM2 = Paediatric Index of Mortality version 2; SEE = standard error of the estimate.BDThe author(s) declare that they have no competing interests.EOD performed data collection, preliminary data analysis and co-wrote the first draft of the manuscript. SMT conceived the study, performed data analysis and co-wrote the first draft of the manuscript. AD participated in the design of the study, derived one of the base deficit formulae and advised on data analysis. JA performed data collection. IAM supervised the project and participated in study design. All authors read and approved the final manuscript.A Word document describing Stewart's physiochemical approach to acid–base balance.Click here for file


A major marketing technique used by pharmaceutical companies is direct-to-physician marketing. This form of marketing frequently employs promotional marketing brochures, based on clinical research, which may influence how a physician prescribes medicines. This study's objective was to investigate whether or not the information in promotional brochures presented to physicians by pharmaceutical representatives is accurate, consistent, and valid with respect to the actual studies upon which the promotional brochures are based.Physicians in five clinics were asked to consecutively collect pharmaceutical promotional brochures and to send them all to a centralized location. The brochures for any class of medication were collected on a continuous basis until 20 distinct promotional brochures were received by a central location. Once the brochure was received, the corresponding original study was obtained. Two blinded reviewers performed an evidence-based review of the article, comparing data that was printed on the brochure to what was found in the original study.Among the 20 studies, 75% of the studies were found to be valid, 80% were funded by the pharmaceutical company, 60% of the studies and the corresponding brochures presented patient-oriented outcomes, and 40% were compared to another treatment regimen. Of the 19 brochures that presented the data as graphs, 4 brochures presented a relative risk reduction while only 1 brochure presented an absolute risk reduction. 15% of the promotional marketing brochures presented data that was different from what was in the original published study.Given the present findings, physicians should be cautious about drawing conclusions regarding a medication based on the marketing brochures provided by pharmaceutical companies. The pharmaceutical industry spent over $11 billion in pharmaceutical marketing, excluding medication samples, in 2004, with more than $7 billion directed to clinicians . Such maThere is an ongoing debate on the effects of pharmaceutical marketing on health care delivery. For example, while diuretic and beta-blocker medications are considered the first line antihypertensive therapies (both medications being relatively inexpensive and available in generic forms), extensive marketing campaigns by pharmaceutical companies to promote calcium channel blockers (CCBs) resulted in the largest growth in revenues for this class of anti-hypertension drugs during the 1990s . The perDirect-to-physician (DTP) marketing is one important facet of the promotion of pharmaceuticals. DTP includes verbal in-office presentations which are usually accompanied by promotional advertising brochures, free medication samples, and possibly gifts such as meals or other promotional items. These in-office presentations by pharmaceutical representatives (PRs) are an important method of promotion for pharmaceutical companies. Their impact is undeniable, as PRs are most frequently reported by physicians as their source of primary information about new medications . While dThere is concern about the influence of DTP marketing on physician prescribing practices and its consequences, such as the physician's ethical obligation to the patient and health care costs. Studies have repeatedly shown that pharmaceutical promotion influences physician behavior ,12-14. TWhile studies have investigated the accuracy and influence of promotional information provided by PRs to physicians, -14,16-21a priori because no studies were available to calculate an accurate sample size. The underlying cited publication on the brochure in support of the identified promotional claim was obtained for review. Two reviewers used a standardized form to collect data and perform an evidence-based review of the article, comparing data that were presented on the brochure to those found in the underlying study. Both reviewers were initially blinded to each other's review. The validity of an article was determined by a series of six questions that addressed evidence-based principles   were calThe characteristics of the 20 brochures are presented in Table A value of .41 was calculated for Cohen's kappa, and the PABAK value was .60, indicating a moderate level of agreement between the two reviewers . The cruSeventy five percent of the reviewed articles were considered valid. This supports findings from a previous study that assessed whether the quality of a study was influenced by sponsorship . The autA large majority of the brochures presented data that were based on underlying studies funded by the pharmaceutical company. However, a recent study by Cooper and Schriger found that only 58% of the original research cited in advertisements was sponsored by a pharmaceutical company or had a company affiliated author . This deAlthough a majority of the outcomes in the reviewed studies were patient-oriented, a substantial number were not. It is important to present data that reflects the outcome of the patient, such as mortality, quality of life changes, and symptom reduction. Thus, while a CCB may reduce blood pressure just as effectively as a diuretic, patient outcomes  may prevail in the long term ,42.While 95% of the brochures presented a visual graphic, only 1 brochure presented an absolute risk reduction and 4 brochures presented relative risk reductions. This "framing effect" may mislead physicians to conclude that a large difference in outcome occurs with the use of the promoted medication or therapy . MoreoveA minority of the studies (eight of 20) compared the medication to another medication or treatment strategy. When a medication has already been approved by the Food and Drug Administration, subsequent studies that compare that medication to placebo have limited value to practicing physicians, as efficacy has already been established. It is of more value to physicians for studies to compare new medications to generic and inexpensive medications that are currently used in practice to determine if a change in disease management is indicated.Although three of the 20 brochures presented data that were different from the original study, the differences were small and most likely would not have affected the clinician's prescribing behavior.This study had several limitations. Only two blinded reviewers were available and any discrepancies were resolved by consensus rather than adjudication by a third reviewer. The method in which the reviewers deemed the underlying studies valid has not been validated in previous studies. The sample size may have been too small to detect clinically important findings. Because brochures were collected only in selected family medicine clinics in one city, these findings may not be generalized to other specialties, settings , or geographic regions. Only brochures were studied, therefore these findings may not be generalized to other types of promotional products or advertisements, such as journal advertisements. We believe that successful drug marketing to physicians entails more than simply presenting a brochure to the physician and this study only assessed one aspect of drug marketing tactics. There may be seasonal variations in drug marketing that further limit generalizability, although this may have been mitigated to some degree by the wide range of drugs included in the study.Given the present findings, physicians should be cautious about drawing conclusions based on data presented on brochures provided by pharmaceutical companies. It would be prudent for physicians to review the original study prior to changing prescribing behavior based on promotional brochures only. Further, physicians should be familiar with and utilize the principles of evidence-based medicine in assessing the validity of published studies. As our study was a descriptive cross-sectional study with the primary purpose of using a semi-objective methodology to review pharmaceutical brochures and their corresponding published studies, future research needs to determine not only how the promotional brochure plays in overall DTP marketing and how pharmaceutical brochures affect physician prescribing behaviors, but also if patient outcomes are associated with such changes.The author(s) declare that they have no competing interests.RC, JCL, and LGT designed and implemented the study. RC and JCL reviewed the brochures and their underlying studies to assess for validity. RC, JCL, and LGT conducted the analyses and drafted the manuscript.The pre-publication history for this paper can be accessed here:


The new findings raise the possibility that pregnant women who inadvertently eat shellfish tainted with low levels of domoic acid may put their unborn children at risk for lifelong behavioral consequences, says Edward D. Levin, a professor of psychiatry at Duke University Medical Center in Durham, North Carolina. Levin coauthored the study, which is described in the September/ October 2005 issue of Neurotoxicology and Teratology.Domoic acid, a naturally occurring marine toxin, causes acute symptoms of diarrhea, vomiting, seizures, and memory loss in people sickened by eating contaminated shellfish. Now a recent study reveals that exposure to even tiny amounts of domoic acid Karenia brevis and other dinoflagellates signal toxic blooms. However, no distinctive color characterizes toxic blooms elsewhere, such as those along the Oregon coast caused by Pseudonitzschia, a phytoplankton that generates domoic acid. “The first warning sign we get for these toxic blooms is when domoic acid shows up in routine testing of shellfish,” says Peter Strutton, a biological oceanographer at Oregon State University in Corvallis. Thus, it’s possible that shellfish with potentially dangerous levels of domoic acid are being harvested and consumed .Harmful algal blooms that produce domoic acid are increasing, possibly due to warming ocean waters and human impacts such as farm and sewage runoff. Shellfish take in the toxin when they filter water and incorporate it into their tissues. Along the Florida coast, pigments in the “red tides” caused by In the Levin study, the researchers injected pregnant rats with single doses of 0.3, 0.6, or 1.2 milligrams of domoic acid per kilogram body weight at the end of the second trimester. The highest dose was in the low end of the range known to cause acute illness in rats.During adolescence and adulthood, the offspring underwent a battery of behavioral tests. In a radial-arm maze, which looks like a wagon wheel without its rim, the rats searched for sugary cereal at the ends of arms extending from a central hub. Once eaten, the cereal is not replaced, and the rats must remember which arms they’ve already explored. This test therefore measures working memory. The females performed the same regardless of dose, while the males performed progressively worse as the dose increased .Next Levin gave the rats low doses of scopolamine, a drug that causes amnesia and memory impairment. Slightly stressing the brain with a low dose of scopolamine helped to uncover subtle neurological defects caused by domoic acid. Compared to controls, rats exposed to domoic acid had greater memory loss following administration of scopolamine, with the highest-dose group performing the worst. “Animals can normally deal with a low dose of scopolamine,” says Levin, “unless there’s prior neurotoxic damage that adversely affects the brain.”Now researchers wonder whether domoic acid may negatively affect unborn children even at levels that do not cause symptoms in expectant mothers. The U.S. Food and Drug Administration based its current limits for domoic acid in shellfish on levels that are assumed to be safe for adults. “We may need to re-evaluate the monitoring of waters and seafood to make sure that the most sensitive members of the population are protected from toxic exposure to domoic acid,” says Levin. However, he adds, it’s important to ensure that fisherman are not unnecessarily cut off from their livelihood and that people are not deprived of the nutritional benefits of uncontaminated seafood.Strutton and Michelle Wood at the University of Oregon in Eugene are developing a new tool to improve early surveillance of toxic blooms. They are combining satellite data on physical attributes of the ocean such as water color and surface temperature to identify early markers for toxic blooms. In collaboration with the National Oceanic and Atmospheric Administration’s CoastWatch program, they plan to develop products for coastal managers such as charts of conditions that raise the risk of domoic acid poisoning of shellfish. CoastWatch also plans to post satellite maps of regions where blooms exist or are developing on its website. “It will warn [managers] to ramp up their shoreline sampling of shellfish beds,” Strutton says.


Apis mellifera constitutes a standard model for the study of appetitive learning and memory since it was shown, almost a century ago, that bees learn to associate different sensory cues with a reward of sugar solution. However, up to now, no study has explored aversive learning in bees in such a way that simultaneous access to its neural bases is granted. Using odorants paired with electric shocks, we conditioned the sting extension reflex, which is exhibited by harnessed bees when subjected to a noxious stimulation. We show that this response can be conditioned so that bees learn to extend their sting in response to the odorant previously punished. Bees also learn to extend the proboscis to one odorant paired with sugar solution and the sting to a different odorant paired with electric shock, thus showing that they can master both appetitive and aversive associations simultaneously. Responding to the appropriate odorant with the appropriate response is possible because two different biogenic amines, octopamine and dopamine subserve appetitive and aversive reinforcement, respectively. While octopamine has been previously shown to substitute for appetitive reinforcement, we demonstrate that blocking of dopaminergic, but not octopaminergic, receptors suppresses aversive learning. Therefore, aversive learning in honeybees can now be accessed both at the behavioral and neural levels, thus opening new research avenues for understanding basic mechanisms of learning and memory.Invertebrates have contributed greatly to our understanding of associative learning because they allow learning protocols to be combined with experimental access to the nervous system. The honeybee  Apis mellifera) constitute a traditional invertebrate model for the study of associative learning at the behavioral, cellular and molecular levels Associative learning allows extracting the logical structure of the world as it enables making predictions about stimuli and their potential outcomes. Honeybees , which is a defensive response to potentially noxious stimuli Honeybees were fixed individually on a metallic holder so that they built a bridge between two brass plates through which a 2 sec mild electric shock (7.5 V) was delivered by a stimulator (60 Hz-AC current) . Bees tr5,190 = 8.46, p<0.0001). Bees in the unpaired group (n = 39) showed no significant change in responsiveness to the unpaired odorant  during trials, thus demonstrating that the increase in SER observed in the paired group was due to associative learning and not to the simple experience with the odorant and the shock, independent of their temporal sequence. The performance of both groups differed significantly over all trials  and the evolution of responses during trials was also different between groups, as shown by the significant interaction . One hour after conditioning, bees of the paired group still remembered the conditioned odorant while bees of the unpaired did not respond to the odorant  was 3 sec and the intertrial interval was 10 min. We compared the performance of these bees to that of bees trained with explicitly unpaired presentations of odorant and shock. Bees of the unpaired group experienced 6 odorant and 6 shock presentations that were temporally dissociated. In all cases we recorded SER to the odorant. Each group was subdivided into two subgroups trained either with 1-hexanol or with eugenol. One hour after the last trial bees were presented with the conditioned odorant alone in a retention test. The two paired subgroups  did not differ significantly and were therefore pooled, as were the two unpaired subgroups. Over successive trials, bees from the paired group (n = 38) significantly increased their response to the odorant that preceded the electric shock . Thus, olfactory conditioning of SER is truly associative and does not depend on the simple exposure to the training stimuli, independent of their outcome.To confirm the associative nature of this learning, we trained bees to extend their sting to an odorant paired with an electric shock and not to respond to a non-reinforced odorant . This procedure is a typical within-subject control in studies of associative learning. Bees were conditioned during 6 reinforced and 6 non-reinforced trials, presented in a pseudo-random sequence with 10 min inter-trial intervals. For one group of bees, eugenol was the reinforced odorant and 1-hexanol the non-reinforced odorant; for a second group the contingencies were inversed. Because there were no significant differences between both groups, data were pooled (n = 48). The resulting learning curves  show tha1,158 = 27.33, p<0.0001) whereas they responded significantly with a PER to the odorant associated with sucrose but not to that associated with electric shock . As a result of training, they exhibited the appropriate response to the appropriate odorant. One hour after the last conditioning trial, bees still responded correctly to the odorants even in the absence of punishment or reward .We then asked whether or not appetitive and aversive olfactory learning can be mastered simultaneously by honeybees. We trained bees to discriminate 1-hexanol and 1-nonanol, one of which was paired with electric shock during 6 trials and the other with sucrose solution 50% (weight/weight) also during 6 trials (SER-PER group). Trials were spaced by 10 min and odorants were presented in a pseudo-random sequence. Data from both conditioned groups (1-hexanol/shock vs. 1-nonanol/sucrose and 1-hexanol/sucrose vs. 1-nonanol/shock) could be pooled (n = 80). The resulting performance is presented in both the aversive and the appetitive odorants (‘doubles’), to the appetitive odorant alone (‘PER only’) and to the aversive odorant alone (‘SER only’). As bees were naïve for both odorants in the first trial, we included only responses from trial 2 to 6 in our analysis. Only the number of bees responding correctly to the aversive and to the appetitive odorant (‘doubles’) increased significantly from trial 2 to 6 . Neither the number of bees responding correctly only to the odorant paired with sucrose  nor the number of bees responding correctly only to the odorant paired with electric shock  varied significantly. For these bees, ‘PER only’ and ‘SER only’, the significant increase occurred between the first and second trial. Thus, at the end of conditioning, 13.75% of bees (11 out of 80 bees) mastered both associations. In the retention tests, the proportion of ‘doubles’ responding correctly to both the aversive and the appetitive odorant was even higher , thus suggesting that additional trials and/or time improves the performance of bees in this double task. The proportion of ‘PER only’ bees in the retention test was 23.75% (19 out of 80 bees) and that of ‘SER only’ bees 27.5% (22 out of 80 bees).To show that bees did indeed learn to master both appetitive and aversive associations simultaneously, we focused on individual performances and quantified the number of bees responding correctly to 1,158 = 31.41, p<0.0001). Similarly, bees learned to discriminate one odorant rewarded with sucrose solution from a non-rewarded odorant so that they extended their proboscis to the rewarded but not to the non-rewarded odorant . One hour after the last conditioning trial, bees in each group responded correctly to the odorants even in the absence of punishment  or of reward . We then compared the performance of these groups  and retention . However, learning both associations resulted in a lower performance for appetitive PER conditioning. Bees in the SER-PER group reached lower levels of PER acquisition and retention compared to the PER group which did not experience electric shocks . Although aversive learning seems to interfere with appetitive learning, lower PER performances in the SER-PER group may simply result from having experienced six electric shocks that would lower the general responsiveness of the bees without implying necessarily interferences between different forms of associative learning.In parallel to the double-task discrimination shown in e groups  to that e groups . Learnine groups  and the e groups , both foTo test this hypothesis we performed an additional control experiment in which we explicitly tested the effect of the electric shock on appetitive PER conditioning. We trained bees to respond to an odorant, either 1-hexanol or 1-nonanol, paired with sucrose solution, and interspersed the shock alone during training. Thus, bees experienced six paired presentations of odorant and sugar reward and six electric shocks following the same pseudorandom sequence as in the previous experiments (‘shock group’). In a parallel group, the same procedure was followed but the shock was substituted by a placement trial in which bees were simply placed in the conditioning setup without receiving any punishment (‘placement group’). One hour after training, bees of both groups were presented in a retention test with the trained odorant and with a novel odorant (1-hexanol for bees trained with 1-nonanol and vice versa) in order to assess the specificity of olfactory memory. As in the previous experiments, trials were spaced by 10 min. The odorants, sucrose concentration and shock intensity were the same as before.5,390 = 33.18, p<0.0001). The acquisition was similar in both groups  and no significant interaction was found . During the retention test, bees of both groups behaved similarly  as they responded significantly more to the learned odorant than to the novel odorant , thus showing that the shocks experienced during training did not affect the appetitive memory one hour after conditioning. These results show, therefore, that the electric shock did not interfere with appetitive olfactory learning. We conclude from these experiments, that while the formation of an appetitive association does not interfere with aversive conditioning, the formation of an aversive odorant-shock association  induces a performance decrease during appetitive conditioning. However, the fact that some bees manage to learn both associations simultaneously supports the notion that appetitive and aversive olfactory learning are mediated by relatively independent neural systems  dedicated to the processing of appetitive and aversive associations.For each group, ‘shock’ and ‘placement’, the two subgroups conditioned to either 1-hexanol or 1-nonanol yielded similar results and were therefore pooled . Both groups of bees learned to respond to the rewarded odorant during training, irrespective of the presence or absence of electric shocks during conditioning  nor for retention . We therefore analyzed the performance of drug-injected groups and compared it to that of Ringer-injected groups.Looking at the neural bases of appetitive and aversive olfactory learning and studying whether the appetitive and aversive unconditioned stimuli used are mediated by different neurotransmitters may help elucidate the relative independence of appetitive and aversive modules. We took advantage of the fact that in our aversive learning protocol bees are immobilized, thus allowing simultaneous access to behavioral performances and to the nervous system. We tested whether octopamine and dopamine, two catecholamines that have been related respectively to appetitive and aversive olfactory learning in fruit flies 1,39 = 67.49, p<0.0001). One hour later, they still remembered the aversive association and extended their sting in response to the previously punished odorant . Octopaminergic antagonists (mianserine or epinastine) did not affect performance at any of the concentrations used in these experiments . Retention tests also showed significant discrimination . An apparent increase of retention was found in bees injected with epinastine with respect to Ringer-injected bees . Consequently, they did not show discrimination in the tests performed one hour later . As an example, Dopaminergic antagonists (fluphenazine and flupentixol) had a dramatic effect on aversive olfactory learning. For both concentrations of flupentixol tested, and contrarily to controls, bees did not learn to discriminate between odorants . One hour later, they still remembered the difference . In vertebrates, dopaminergic receptors are generally classified in two main families, the D1-like and D2-like receptors 1,39 = 2.77, NS) but not at the high dose  (n = 40 in both cases). SCH23390 had no effect on retention as bees responded significantly more to the odorant previously paired with the electric shock and did not differ from control bees . SCH23390-injected bees failed to show evidence of acquisition at the low dose  nor did they show any retention one hour later . AmDOP receptors could contribute differently to aversive learning in bees.By contrast, spiperone-injected bees were significantly impaired in both learning and retention at both doses used (n = 40 in both cases) as they did not learn the difference between reinforced and non-reinforced odorants  learn to master both appetitive and aversive olfactory learning simultaneously, so that they exhibit the appropriate response, PER or SER, to the appropriate odorant. Bees trained in such a way exhibit, however, lower scores for appetitive learning than bees trained to discriminate two odorants in the appetitive modality (one odorant rewarded with sucrose solutions vs. a different odorant non-rewarded). This difference was not due to a detrimental effect of the electric shock as shown by the fact that appetitive olfactory learning was unaffected by interspersing shocks among odorant-sugar trials. The lower performance in appetitive conditioning (PER) found in bees trained with the double discrimination task  could beDrosophila brain visual stimuli associated with aversive gustatory USs To determine whether the relative independence of appetitive and aversive learning is supported by independent modules corresponding to separate neural systems dedicated to the processing of the different unconditioned stimuli, we took advantage of the possibility of combining aversive conditioning and neuropharmacological tools. Octopamine has been strongly implicated in appetitive olfactory learning in bees and octopamine injections in the brain can substitute for sucrose reward and induce olfactory learning AmDOP receptors in aversive learning and memory in bees. Studying these questions will require, however, the development of specific pharmacological antagonists or the use of RNAi techniques already applied successfully in honeybees We also showed that differences in learning and retention can be found on the basis of their susceptibility to blocking by dopaminergic ligands. The ligands used are well-known vertebrate D1-like and D2-like receptor antagonists. Although the specificity of these antagonists is not granted for insect receptors, the fact that the D1-like antagonist SCH23390 was less effective than the D2-like antagonist spiperone, which suppressed aversive learning and memory, raises the question of the differential involvement of For honeybees, stinging can eventually lead to death because the sting may remain attached to the stung surface Finally, we discuss the pertinence of the terminology used to characterize our conditioning protocol. We have defined the olfactory conditioning of the sting extension reflex as a case of aversive learning. Contrarily to other cases of aversive learning, however, the consequence of SER conditioning is neither an avoidance response towards the stimulus (here a given odorant) predicting the noxious unconditioned stimulus (here the shock) nor a response inhibition. In traditional protocols of aversive learning, individuals learn to actively avoid a noxious stimulus  was always 10 min. Retention tests were performed 1 h after the last conditioning trial and consisted of presenting odorant stimuli without punishment. In order to ensure an ITI of 10 min, groups of 10 bees were trained one after the other. In this way, 10 min were required for the bees to complete each trial and to move to the next trial. Several conditioning apparatuses were available to run several groups in parallel if necessary.In the paired group , conditiExperiments combining appetitive PER and aversive SER conditioning  were perIn all experiments, US responses were measured before and after conditioning or retention test. Only bees which consistently reacted to the electric shock were taken into consideration. This is particularly important when testing the effect of drug injection (see below) as it precludes the possibility that the analysis reflects the action of a given antagonist on a pathway controlling only the sting extension reflex, irrespective of the possible association with the paired odorant.2 4 mM, CaCl2 5 mM, HEPES 10 mM, glucose 25 mM, sucrose 160 mM]. Drugs or Ringer alone (control) were injected into the brain through the median ocellar tract. In addition to normal preparation, the head of each harnessed bee was fastened to the holder with a small drop of wax to avoid movements. A Harvard GC 100-10 microelectrode filled with the drug to be injected was connected to an IM 300 Narishige microinjector and used to deliver 10×20 nl into the brain. Volumes injected were calibrated before and after injection by means of a Malassez cell. Injections were performed 30 min before conditioning, as earlier experiments (reviewed in ref. 42) showed that pharmacological injections of catecholamines and their inhibitors are effective approximately 30 min after drug application. Octopaminergic  and dopaminergic receptor antagonists   were used at µm and mM doses , as it constitutes an appropriate measure of the discrimination reached at the end of conditioning, and for the retention test. Comparisons between groups were done using a Mann-Whitney test. Bonferroni corrections were used to adjust the alpha levels of comparisons involving groups performed in parallel.During the experiments, we recorded the response to the presented odorant, i.e. whether bees extended their sting after the onset of the odorant and before the presentation of the electric shock in the case of reinforced trials so that conditioned responses were recorded. Multiple responses during odorant presentation were counted as a single SER. The percentage of SER recorded during acquisition was used to plot acquisition curves. Similar procedure was followed for PER in the case of appetitive learning. To analyze the variation of performance during trials, we used analyses of variance (ANOVAs) for repeated measurements both for between-group and for within-group comparisons. Monte Carlo studies have shown that it is permissible to use ANOVA on dichotomous data only under controlled conditions


Previous studies have proposed correlation between variants of the cerebral arterial circle  and some cerebrovascular diseases. Differences in the incidence of these diseases in different populations have also been investigated. The study of variations in the anatomy of the cerebral arterial circle may partially explain differences in the incidence of some of the cerebrovascular diseases in different ethnic or racial groups.While many studies have investigated the variations in the anatomy of each segment of the cerebral arterial circle, few have addressed the variants of the cerebral arterial circle as a whole. Similarly, the frequency of occurrence of such variants in different ethnic or racial groups has not been compared.102 brains of recently deceased Iranian males were dissected, in order to observe variations in the anatomy of the cerebral arterial circle. The dissection process was recorded on film and digitized. One resized picture from each dissection, showing complete circle has been made available online. The variations of the circle as whole and segmental variations were compared with previous studies.On the whole, the frequencies of the different variants of the entire cerebral arterial circle and segmental variations were comparable with previous studies.More specifically variants with uni- and bilateral hypoplasia of posterior communicating arteries were the most common in our study, similar to the previous works. No hypoplasia of the precommunicating part of the left anterior cerebral artery (A1), aplasia of A1 or the precommunicating part of the posterior cerebral artery (P1) was seen. In 3% both right and left posterior communcating arteries were absent.The anatomical variations found in the cerebral arterial circle of the Iranian males in the current study were not significantly different to those of more diverse populations reported in the literature. While taking into account potential confounding factors, the authors conclude that based on available studies, there is no evidence suggesting that the distributions of the variations of cerebral arterial circle differ in different populations. The cerebral arterial circle  was described by Thomas Willis in 1664, but in majority of cases there are some variations from the original definition.Previous studies have proposed correlation between variants of the CAC and some cerebrovascular diseases -4 and alThe CAC is the main route for collateral blood flow in severe occlusive diseases of the internal carotid artery. Those patients with variants of the circle with efficient collateral circulation have a lower risk of transient ischemic attack (TIA) and stroke than that of patients without such collaterals,2. The iThere are studies of racial variations in location and risk of intracerebral hemorrhage which can not be explained completely by known risk factors .Other studies have found a correlation between cerebral aneurysms and certain variations of the CAC . FinallyWhether the distribution of variations in different populations is similar or not may help us in clarifying the potential role of variations of the CAC in some of the cerebrovascular diseases.Many previous studies have investigated the variations of each segment of the CAC ,14-21 buIn order to clarify the above questions, the authors reviewed many series. Four studies were identified, which presented the variants of the entire CAC in detail , the precommunicating part of the anterior cerebral arteries (A1), the precommunicating part of the posterior cerebral arteries (P1) and the posterior communicating arteries (PcoA).The authors utilized two means of comparison with previous autopsy studies. Firstly, the incomplete circles were excluded .Using computer software Osiris, downloadable for free from world wide web , The extIn this study's classification, segments without hypoplasia were called normal segments. figure  Type 1 –Since the above classification resulted in many cells with zero values, the use of common statistical methods for comparisons was not appropriate.As several vessels were thrombosed and many were collapsed due to extravasations, measurements of the external diameters sometimes had to be estimated. In the case of fenestration of the AcoA, the diameter of the whole fenestrated vessel was measured.In order to study the interobserver variability in the measurements, we performed paired T-test between first and second measurements done on thirteen cases.Segmental variations were also studied and compared with similar studies in the past.102 fresh cadavers of Iranian male subjects aged between 15 and 75 years (mean 38 ± 15 years) were dissected.Paired T-test showed no significant statistical difference between first and second measurements done on 13 cases.The frequencies of different types of the CAC as a whole (92 complete circles) are shown in Table As can be seen in figure The diameters of 26% of P1s on the right side and 28% on the left side were smaller than the PcoAs on the same side.Frequencies of segmental variations using all 102 cerebral arterial circles are shown in the Table No hypoplasia of left A1, aplasia of A1 or P1 segments was seen. Hypoplasia of right PcoA in 16 cases (16%) and left PcoA in 11 cases (11%) were seen. In 34 cases (33%) both right and left PcoAs were hypoplastic. Aplasia of AcoA in 1 case (1%), right PcoA in 4 cases (4%) and left PcoA in 3 case (3%) was seen. In 3 cases (3%) both right and left PcoAs were absent. In those with no anastomosis between the carotid and vertebrobasilar systems, we could not find any primitive trigeminal artery, but we did not investigate the presence of other primitive caroticobasilar anastomoses.Although this was not our primary objective, we found one aneurysm of the PcoA-P1 junction.In this study the anatomical variants of the entire CAC were classified and compared with results of four previous similar studies. The segmental variations were also compared with previous published data.Before discussing the main findings and comparisons, we would like to address the main limitations of the study.There are some limitations which apply generally to all such similar studies. One of them is potential minor changes in the diameter of the vessels during time. In other words, the so-called hypoplastic vessels may change to vessels with diameters larger than 1 mm during time. What we have studied is a snapshot of the circle with arbitrary definition of one millimeter as the criterion for differentiation of hypoplastic segments.Fisher  and manyWe can never be sure that our postmortem measurements are similar to those in real life. Some technical considerations may minimize the potential sources of error.This problem may raise some concerns about reliability of our results. Our definition of hypoplasia is based on our measurements done on the images of the vessels using computer software. The measured diameters of collapsed vessels may be less precise than or different from those measured after removal of the circles and compressing them between the glass plates, formalin fixation or using injection techniques. Since other sources for measurement were not available  and we could not find any published data that had used such a method of measurement in medicine, we could not comment on the extent of errors. On the other hand, we have not used the absolute values of the measurements and only used them for classifications. This may minimize the extent of error and make the method acceptable for this purpose. Utilization of advanced imaging technology like MR or CT angiograms and related computer software can provide us with more precise measurements. We think that classifying the segments according to hypoplastic and normal does not seem to be realistic. Future studies should look at the circle as a dynamic structure and use classifications based on absolute values of the luminal diameters and flow of the segments. Several potential confounding factors should be considered for analysis of the results and comparisons.We selected our cases from autopsies done on subjects in whom the apparent cause of death was either "natural" or "traumatic". This is similar to El Khamlichi et al  who had We restricted our cases to male subjects. El Khamlichi et al  had studRiggs and Rupp , Fisher Race may be one of the factors that may account for the differences in the distribution of the variations. Our study population was Iranian. We had excluded other races from our study, in order to control the potential confounding factor. None of the previous studies had discussed the racial homogeneity among their subjects.Even though we have tried to use similar classifications, there are still some uncertainties about similarity of definitions and methods of measurement. The definition of hypoplasia has not been quantitatively defined in works of Riggs and Rupp  and of LWe addressed the issue of interobserver variability by doing a limited study on repeated measurements. Although the study showed that there were no significant statistical differences between repeated measurements, there were still a few controversial cases. These cases were classified according to the consensus among the authors.Putting aside the internal carotid segments of the CAC and classifying the segments into hypoplastic or not, we can theoretically have 64 variants of the entire circle, irrespective of the side. We have seen less than 31 variants in our and previous studies.As is seen in Table Although Type 3 (hypoplastic AcoA) was seen in 10% of Riggs and Rupp  cases anType 17  was seen in less than 3% of all studies except Fisher's. Fisher saw Type 17 in 11% of cases .There are some similarities and differences between previous studies and ours. Considering the above- mentioned potential reasons for differences, we could not reach the conclusion that the distributions of variations of the entire CAC are different among different populations. As mentioned previously, that there were many cells with zero values, therefore, we can not support our findings with common statistical methods.The classification used by Lazorthes et al  and thisBased on the above mentioned comparisons and classifications, we could not conclude that the distributions of variations of the entire CAC are different in different populations. Obviously this conclusion might be untrue using other classifications.As is seen in Table The fetal form of the posterior cerebral artery . Piganiol et al  observedThe frequencies of the segmental variations of the CAC other than bilateral absence of PcoAs were within the range of previous studies.Variation of the CAC has been the subject of many theories. Lazorthes et al  suggesteHillen , in his According to Stehbens , topograVan Overbeeke and colleagues , examiniMilenkovic et al  based onVasovic et al  believe From the evolutionary standpoint, it is noteworthy that variations of the cerebral arteries seem to be equally common in humans as well as animals . AlthougThe online availability of some of the digital images may be considered an advantage of this study and helps repetition of some of the measurements and comparisons. Resizing of the images may have decreased the quality of some images. A few of the images are not of good quality, but since they could be classified, we finally decided to include them.As discussed before, the results should be considered cautiously and need further verifications. Advanced imaging techniques can help us significantly in this regard.In this study, we presented our experience and reviewed many other series, trying to figure out whether the distribution of the variations of the CAC is similar among different populations or not.Considering potential confounding factors, we conclude that we have no evidence that the distributions of the variations of CAC differ in different populations.The results should be interpreted cautiously. In order to verify these findings, further studies are needed. We recommend using new imaging modalities and classifications of variants of the entire CAC based on quantitative measurements of the luminal diameters and flow of the segments.The author(s) declare that they have no competing interests.BE carried out the data extraction, performed the analysis and drafted the manuscript. MD and SA carried out the data extraction and participated in analysis. MG, BN and EK supervised the study and participated in its coordination. The figures were drawn by MD. All authors read and approved the final manuscript.The pre-publication history for this paper can be accessed here:


The role of the Beta2-adrenoceptor (beta2-AR) Gln27Glu polymorphism in the manifestation of cardiovascular diseases is still unclear.In the present study, we evaluated the potential relevance of the c.79 C>G (p.Gln27Glu) polymorphism of this receptor gene for coronary artery disease (CAD) and its associated risk factors in Saudi Arabs. Genotyping was performed by PCR using the confronting two-pair primer (PCR-CTPP) method.p = < .001) and G/G (p = .005) genotypes are significantly associated with CAD, when compared to the CON group. In addition, C/G (p = < .001) and G/G (p = < .001) were significantly associated with CAD, when compared to the BD group. Furthermore, stepwise logistic regression showed that the genotype [C/G (p < .001) and G/G (p < .001)] increase the risk of CAD.In the general population group (BD) (n = 895), 68.5% were homozygous wild-type C/C, 28.3% were heterozygous C/G and 3.2% were homozygous mutant G/G. Among the CAD patients (n = 773), 50.6% were homozygous wild-type C/C, 43.6% were heterozygous C/G and 5.8% were homozygous mutant G/G, while in the angiographed control group (CON) (n = 528), 71.8% were C/C, 24.4% C/G and 3.8% G/G genotypes. These results indicate that both the C/G (These results shows that the Gln27Glu genotypes (homo- or heterozygous) of the beta2-AR may be independent predictors of severe CAD. Hence, mutations in the gene coding for this protein are likely to have a great impact on cardiovascular reactivity and function. The gene is highly polymorphic in humans, the most common of which are the three polymorphisms Glu27, Gly16 and Thr164 that have been invariably associated with different cardiac risk factors. Besides, there is conflicting data in the literature, specifically with regard to the relevance of the Glu27 polymorphisms for the manifestation of different cardiovascular diseases and in different populations. For example, while some investigators found an association between this polymorphism and obesity in French population [2-AR gene and CAD in Arabs. Therefore, these genotypes (C/G or G/G) could be used as a predictor of severe CAD at least in our population.The β disease . Apart fpulation -5. Besidpulation , but notpulation -12, bloopulation , obesityThree groups of Saudi individuals were recruited for the present study. The patient group comprised 773 candidates  of Saudi Arabian descent with angiographically documented CAD (CAD group). The inclusion criteria for severe CAD involved, among others, the presence of angiographically established narrowing of the coronary vessels by at least 70%. A second group of 528 individuals  undergoing surgery for heart valvular diseases and those who reported with chest pain, but were established to have clear vessels by angiography, were recruited as controls (CON) group. A further group of 895 healthy Saudi individuals  visiting the Blood Donor Clinic at King Faisal Specialist Hospital and Research Centre from January 2003 to November 2004, were recruited as representing the whole spectrum of the Saudi general population (BD group). This study was performed in compliance with the Helsinki declaration and in accordance with the regulations laid down by the Hospital Ethics Committee  and all participants signed an informed consent.Five ml of peripheral blood were collected in EDTA tubes from all participating individuals after obtaining their written consent. DNA was extracted using the PURGENE kit from Gentra Systems , and stored in aliquots at -20°C until required. Serum cholesterol and triglyceride levels were measured as routine in the main Hospital Pathology Laboratory. Triglyceride levels < 1.8 mmol/L and total cholesterol levels < 5.2 mmol/L were considered normal. Diabetic patients either had a known history of diabetes mellitus or were diagnosed according to the American Diabetes Association criteria . Diagnos2-AR was determined by polymerase chain reaction (PCR) procedure using a modified confronting two-pair primer (PCR-CTPP) method described previously [The genotyping for the c.79 C>G (p.Gln27Glu) polymorphism of βeviously . As a qu2-AR genotype on CAD status, incorporating other variables, such as coronary risk factors, into the model. A two-tailed p value < .05 was considered statistically significant. All analyses were performed using the SPSS v.10  statistical analysis software.Genotype frequencies among the various groups were compared by Chi-Square test. Multivariable logistic regression was used to study the effect of the βp = < .001), see Table p = < .001), Table In the BD group (n = 895), 613 (68.5%) were homozygous wild-type C/C, 253 (28.3%) were heterozygous C/G and 29 (3.2%) were homozygous mutant G/G. Among the severe CAD patients (n = 773), 391 (50.6%) carried the C/C genotype, 337 (43.6%) were heterozygous C/G and 45 (5.8%) and were homozygous for the G/G genotype. Comparison of the CAD with the BD group, using the C/C wild-type genotype as the reference, both the C/G and G/G genotypes were associated with CAD , see Table p = < .001), Table Among the CON group n = 528), 379 (71.8%) were C/C, 129 (24.4%) were heterozygous C/G and 20 (3.8%) were homozygous for the G/G genotype. Comparison of the CAD and CON groups, using the C/C wild-type genotype as the reference showed that both the C/G and G/G genotypes were associated with CAD , age and diabetes mellitus were associated with CAD, whereas elevated cholesterol (p = .313), elevated triglycerides (p = .220), family history of CAD (p = .395), gender (p = .057), hypertension (p = .894) and myocardial infarction (p = .614) were not, see Table p = < .05) were then put into a stepwise logistic regression in order to study the possible combined effect of the genotypes with other risk factors on CAD manifestation. The only variable retained in the model was the genotype (p = < .001), see Table p = < .001) and G/G (p = < .001) with severe CAD manifestation.Univariate analysis showed that βrol p = .13, elevan p = .89 and myocIt is noteworthy that since the consanguinity rate in the Saudi population is > 65% , it was 2-AR gene we selected 895 Saudi blood donors representing the whole spectrum of the Saudi population. To our knowledge this is the largest group thus far involved in studying the frequency of the β2-AR (Glu27) genotypes. The present study shows that the homozygous C/C genotype of this gene is the most predominant, while the G/G is the least prevalent in the Saudi population. The frequency of the C/G genotype (28.3%) was higher than the frequencies of 15, 16, 10, 13, 26 and 19% observed among Ghanaians, Kenyans, Sudanese, Filipinos, Indians and Chinese populations respectively [For determining the genotype frequencies of the βectively . Howeverectively .The frequency for the G/G genotype was 3.2%, which is almost similar to the rate observed among Ghanaians, Filipinos and Indian populations respectively. However, our rate was lower than the rates of 21, 15, 14 and 11% observed among Scottish, Caucasian USA, Swedish and Sudanese populations respectively .2-AR gene among the different populations.The observed rate of 17% for the Glu27 allele  is similar to the 21% found among the African-Americans , but app2-AR polymorphisms in these diseases has been inconsistent, with some studies pointing to an association and others showing no relationship of these polymorphisms with various cardiovascular disorders. Our observation together with studies by other investigators furnishes support to the notion that the β2-AR Glu27 polymorphism may be invariably important for the manifestation of CAD in different populations.Our results also indicate strong association of both the C/G and G/G genotypes with the manifestation of severe CAD in Arabs. However, further studies in larger populations are required to verify these results. Nonetheless, it is noteworthy that, while this mutation has been associated with various CAD risk factors such as obesity , rheumatIn summary, our results indicate a strong association between the C/G and G/G genotypes and the Glu27 (G) allele with severe CAD in Arabs. Therefore, these genotypes (C/G or G/G) could be used as a predictor of severe CAD at least in this population.2-AR Beta-2-adrenoceptorβBD Blood donorsCAD Coronary artery diseaseChol CholersterolCI Confidence intervalCON Control groupDM Diabetes mellitusFH Family history of CADMI Myocardial infarctionPCR Polymerase chain reactionTG TriglycerdiesThe author(s) declare that they have no competing interests.KKA was in charge of design and analysis of data, OMA performed the technical aspects of the study, PCR and genotyping, GHM performed the statistical analysis and ND was responsible for recruiting patients and overall supervision of the study.The pre-publication history for this paper can be accessed here:


The study design with the smallest bias for causal inference is a perfect randomized clinical trial. Since this design is often not feasible in epidemiologic studies, an important challenge is to model bias properly and take random and systematic variation properly into account. A value for a target parameter might be said to be "incompatible" with the data (under the model used) if the parameter's confidence interval excludes it. However, this "incompatibility" may be due to bias and/or extra-variation.We propose the following way of re-interpreting conventional results. Given a specified focal value for a target parameter , the difference between the focal value and the nearest boundary of the confidence interval for the parameter is calculated. This represents the maximum correction of the interval boundary, for bias and extra-variation, that would still leave the focal value outside the interval, so that the focal value remained "incompatible" with the data. We describe a short example application concerning a meta analysis of air versus pure oxygen resuscitation treatment in newborn infants. Some general guidelines are provided for how to assess the probability that the appropriate correction for a particular study would be greater than this maximum .Although this approach does not yet provide a method, because the latter probability can not be objectively assessed, this paper aims to stimulate the re-interpretation of conventional confidence intervals, and more and better studies of the effects of different biases. AlthouConventional frequentist analyses often yield biased point estimates because they implicitly set all bias parameters (e.g. misclassification probabilities) to zero. Bias also arises from misspecation of models, e.g. the ignoring of covariates or of heterogeneity in individual effects ,5. MoreoIn this paper we use the term "extra-variation" to summarize uncertainties about biases and unmodelled random error. The term "bias" denotes all biases and is defined as the difference between the expectation of the estimator and the true average causal effect. Strictly speaking, we mean "expected bias" because we invoke a Bayesian model that involves priors.There are basically five approaches to address bias and uncertainties about bias :1. Most frequently, only intuitive discussions about the impact of bias are provided. These are often found to be wrong when evaluated quantitatively, for instance, when misclassification is falsely assumed to be non-differential for all individuals ,16.2. Sometimes, a subsample is used to investigate a single source of bias (most often measurement error). Such validation studies, however, are often small, giving large random error, and may also be biased, e.g. due to selection bias .3. In sensitivity analysis several bias scenarios are investigated. Bias parameter values are added to the model and values specified for these. The data are then analysed supposing these values to be true, and the dependence of the results on the assumed values of these unknown bias parameters examined. However, sensitivity analysis often shows only that a great variety of results are possible if bias parameters are chosen accordingly.4. In Monte Carlo sensitivity analysis (MCSA), distributions for the bias parameters are specified. Causal-effect estimates are then sampled by drawing bias parameter values from their distributions and repeating the analysis for each draw .-5, is a priori as probable as a risk ratio of 1.5  excludes 0. If lobs > 0, the MPC (in the lower boundary) simply equals lobs, as we require that lobs- a > 0. This result is often applied intuitively: it simply indicates that, the further the lower boundary is from the null, the more room there is for bias and extra-variation. If uobs < 0, the MPC (in the upper boundary) equals uobs.Suppose the focal value is 0, and that the interval [θ, exceeds some pre-specified non-null focal value, κ, that reflects a relevant threshold for clinical or policy significance. Assuming lobs> κ, the MPC is lobs - κ. Likewise, one may ask whether θ is smaller than a certain κ which corresponds to little harm. Assuming uobs< κ, the MPC is κ - uobs.Alternatively, one may require that the causal effect, θ = log RR, one is often interested in effects of at least a q-fold risk (e.g. q = 2). Here, κ = log(q), so that a must be less than lobs - log(q).If θ is RD, then demonstrating a RD of more than κ (e.g. κ = 0.1) requires that a be less than lobs - κ. RD equals the inverse number needed to treat (NNT), the number of individuals required to prevent (or delay) one adverse event. Although NNT has often been misinterpreted [q is equivalent to specifying that RD > κ = 1/q.If erpreted , this meerpreted . Showinga or b, is less than the maximum correction that would leave the focal value "incompatible" with the data. This probability should be assessed with respect to understood bias. Of course, the result could still be distorted by non-understood or unknown bias. First, one should assess the shift due to bias by looking at studies that have investigated specific biases or global bias. Second, a further correction should be added for the uncertainty about the bias parameter values. The magnitude of this correction depends on the size of the studies from which bias was estimated, the uncertainty about their applicability to the present data and assumptions made about bias in these studies. Third, a correction due to extra random variation should be added. The magnitude of the true shift could be estimated from studies  that have compared, in similar settings, naive interval estimates with estimates obtained using more sophisticated methods. Finally, one would compare the shift, a or b, estimated using the above procedure with the MPC for the focal value.We need to estimate the probability that the true shift, a and b to be estimated based on understood bias. Knowledge of such analyses should enable researchers to improve their assessments about a and b. In the discussion we give more guidance for assessing these shifts.This approach aims to improve the intuitive assessment of bias by relating it to MCSA and Bayesian methods. These procedures allow Davis et al. performed a meta analysis of five clinical trials of 100 % oxygen versus air resuscitation treatment of newborns . ResusciOf the four trials, one was randomized and the care-providers and outcome-assessors were blind to treatment status. The other three studies were only quasi-randomized and without blinding. Three studies allowed backup-therapy with oxygen therapy if air therapy was unsuccessful and one study excluded the individuals who received backup therapy. Davis et al  used a fDavis et al interpret their results as evidence to prefer initial use of air resusciation and to use oxygen as backup if necessary. This is a conclusion about clinical practice: about which treatment to use first given that the other treatment may be available as back-up. If one is interested in the pure efficacy of the competing treatments, i.e., how well they work when used alone, then in the absence of other biases, the RD estimate is likely to be biased towards no effect (i.e. underestimated) because of the availability of back-up oxygen treatment in three studies. The lack of blinding in three of the four studies could have caused over-estimation of RD, as has been found in other studies . PublicaThere are various likely sources of extra-variation in the estimated risk difference. First, there is the uncertainty in the bias parameters: the results of Davis et al. assume these are zero. Second, there is unmodelled random variation, e.g. due to heterogeneity in effect magnitude between the studies. Heterogeneity was not statistically significant but this may be due to low power, and it is known that fixed-effect models underestimate the standard error when there is heterogeneity . Unmodela, is greater than 0.014: no clear preference for either treatment can be inferred.In conclusion, there could be several sources of bias and extra-variation in the results. Information about the magnitude of the bias parameters appears sparse, so large uncertainty about them should be allowed, as well as considerable unmodelled random error. It seems likely that the true, unknown shift, The approach advocated in this paper involves the assessment of the probability that the true shift in the relevant confidence bound is less than the maximum permitted correction. Even for experts this is a very difficult task and experts might disagree substantially in their assessment. From cognitive psychology it is known since the 1970s that people tend to rely on a small number of simple heuristics when faced with decisions based on probability assessments -36. The understood bias. The inference could still be distorted by misunderstood or unknown bias, as well as by residual uncertainty.1. The probability that the maximum permitted correction is sufficient should be specified according to anchor information". In many applications, specific biases have been investigated, for instance, by comparing an instrument with small measurement error with another known to have much larger error. However, such estimates of specific biases and associated uncertainty are themselves error-prone because of potential incomparability of studies with respect to other biases (and random error). Moreover, the same kind of bias might operate differently in different studies. For instance, an instrument with good validity in clinical populations might perform poorly in general populations. Therefore using information from other studies has to be done very carefully because it could cause more harm than good.2. We recommend using information from similar studies or MCSA and Bayesian analyses applied in similar situations. The results from such studies can be used as a crude basis for the assessment of the probability that the MPC is sufficient. In cognitive psychology, information used for an assessment that is taken from similar entities or settings is called "Likewise, cognitive psychology tells us that anchor information can be quite misleading -42 and t3. The shift in the interval boundaries could be directly estimated by MCSA or Bayesian methods. However, these analyses are not easy for non-experts to conduct. We expect, on the other hand, that the more results researchers see from such analyses, the more they would develop an intuitive feeling for the effects that multiple bias and extra variation might have in specific situations. At the very least, such analyses show that one should have much less confident in conventional analyses than is suggested by their confidence intervals.However, given these guidelines there remains much uncertainty and subjectivity in assessing the probability that the MPC is sufficient. Therefore, the way of re-interpreting confidence intervals described in this paper does not yet constitute a method. Further studies on bias are required to provide more objective information and to render the approach more useful. This paper is intended to be just a starting point for thinking about re-interpreting conventional confidence limits.MCSA = Monte Carlo sensitivity analysisCI = confidence intervalRR = risk ratioOR = odds ratiolog = natural logarithmRD = risk differenceNNT = number needed to treatMPC = maximum permitted correctionThe author(s) declare that they have no competing interests.Both MH and SRS have contributed to all aspects of the content.The pre-publication history for this paper can be accessed here: