To explore the contributions of different cognitive and personality components, participants also completed a series of tests respectively assessing general speed, inhibitory control, short-term and working memory, cognitive flexibility, crystallized and fluid intelligence, and personality. Stepwise regression analyses revealed that online idiom comprehension was associated with the participants' working memory, inhibitory control and crystallized verbal intelligence, an association modulated by idiom type. Also personality-related variables (State Anxiety and Openness to Experience) were associated with idiom comprehension, although in marginally significant ways. These results contribute to the renewed interest on how individual variability modulates language comprehension, and for the first time document contributions of individual variability on lexicalized, high frequency multi-word expressions as idioms adding new knowledge to the existing evidence on metaphor and sarcasm.In this exploratory study, we investigated whether and to what extent individual differences in cognitive and personality variables are associated with spoken idiom comprehension in context. Language unimpaired participants were enrolled in a cross-modal lexical decision study in which semantically ambiguous Italian idioms (i.e., strings with both a literal and an idiomatic interpretation as, for instance,  Idioms still represent a challenge to theories of language processing after more than 40 years from the first psycholinguistic study on idiom comprehension , the reader/listener has to decide between competing interpretations favoring the one contextually relevant and inhibiting already processed and actually irrelevant information associated to individual word meanings.Several models have been proposed to account for idiom processing (see below). Notwithstanding, the complex cognitive architecture that stands behind the relatively easy and fast comprehension of familiar idioms has not yet been singled out in full details. Can we take for granted the assumption that the cognitive architecture underlying language comprehension is precisely the same no matter whether the input is literal or idiomatic? The answer is that we do not know yet. There are reasons to suspect a different involvement and/or weight of some cognitive components  when the input is an idiomatic string compared to when it is a matched literal string. In fact, idioms are somewhat “special” in that they behave at the same time as partially compositional string of words (as any other piece of language) and as non-compositional string of words: while the meaning of a sentence is obtained by merging individual word meanings as the message unfolds, in idiomatic sentences what must be merged comprises unitary sequences of co-occurring words  and components of the personality structure (Big Five Observer). Measuring state and trait anxiety may reveal whether or not increased emotional arousal associated to the specific event (the experiment) or to the individual's personality may lead participants to recruit more attentional and processing resources speeding up mental processing . All were Italian native speakers with normal or corrected-to-normal vision, unaware of the aim of the study. This study was carried out in accordance with the recommendations of the “Italian Association of Psychology” (AIP) Ethical Guidelines (Codice Etico: www.aipass.org/node/11560) and with the standard procedures of the University of Modena-Reggio Emilia. All subjects gave written informed consent in accordance with the Declaration of Helsinki.Sixty-four students volunteered to participate in the study  were selected from a list of Italian idioms. Twenty-five university students were asked to rate the familiarity of each idiom on a seven-point scale (from 1: never heard to 7: heard very often) reported under each idiom. At the end of the questionnaire, they went back to the beginning and paraphrase each idiom. Since the task was rather demanding in time, the idioms were split into two questionnaires. We selected 48 idioms highly familiar  and correctly paraphrased by 95% of the participants.Age of acquisition (AoA). Since the AoA of idioms correlates with idiom familiarity and knowledge of the idiomatic meaning , in line with what has been shown in the idiom acquisition literature  and 48 filler fragments of similar length and structure. Twenty-five students were asked to complete each sentence with the first word that came to mind. The idioms that were completed idiomatically with a cloze probability of at least 70% were considered as predictable (P idioms) and those that received <30% of idiomatic completions as unpredictable (UP idioms). To ensure that the completions were not due to literal interpretations, each participant was also asked to provide a meaning paraphrase of each sentence. Participants always provided idiomatic paraphrases. Twenty-eight idioms were selected, 14 predictable and 14 unpredictable, balanced for familiarity and AoA . The figurative meaning was similarly perceived as the dominant one in P and UP idioms.Target selection. To select the idiomatic visual targets, 15 participants were asked to provide a word semantically associated to the idiomatic meaning of the string. The words more frequently associated to each idiom were listed under the idiom string in a written questionnaire. Fifteen students were asked to rate on a 5-point scale the extent to which each word was associated to the figurative meaning of the idiom string . The words with the highest association rating were used as visual targets. The mean association ratings for the targets of P and UP idioms were similar. To control that the selected targets were not associated with the last word of the string, 10 students were presented with the last word of each string and were asked to produce the first associated word that came to mind. None of the experimental targets was generated. The targets for P and UP idioms were balanced for number of characters and syllables, AoA, written frequency, and familiarity . The filler words were all abstract. Each stimulus was visually presented in a randomized order to 15 participants who performed a visual lexical decision task. Lexical decision times were statistically similar for idiomatic and control targets; for idiomatic targets associated to P and UP; and for controls associated to P and UP idioms.Sentential contexts. The 28 idiom strings were inserted in contexts orienting the interpretation toward the figurative sense of the string. The average number of words composing them was 14.89 (SD = 1.03). The idiomatic contexts associated to P and UP idioms had a similar number of words.M = 14.94, SD = 1.22) and similar syntactic structures, but without any figurative expressions, were created. A visual target was paired to each filler sentence: 56 targets were legal non-words and 28 were words, so that we had an equal number of words and non-words within each list. The filler target words had on average the same number of syllables of the experimental words. Fourteen of the filler words were semantically associated to the sentence in order to avoid that participants developed a response set to the experimental trials. All filler words had an abstract meaning.In addition to the 28 experimental sentences, 84 filler sentences of the same length  and, interspersed among the experimental sentences, the 84 filler sentences. The experiment was preceded by a practice session formed by 10 different literal sentences, followed by a word in half of the cases and by a legal non-word in the other half, that were not further analyzed.The stimuli were presented on an IBM-compatible personal computer using DMDX software  each time a star appeared on the screen. The randomly distributed inter-trail intervals were 600, 900, 1,200, and 1,500 ms. There was a practice session of 10 trials followed by 80 experimental trials. We recorded the response times between the onset of the stimulus and the response. Response times exceeding 1,500 ms were considered as errors.Inhibitory control: (1) Go-No Go task. Participants were asked to press the space bar as fast and accurately as possible when a triangle (Go stimulus) appeared on the screen and refrain from responding when a circle appeared (No-Go stimulus). There were 200 trials, of which 160 were Go trials and 40 No-Go trials, with randomly distributed inter-trial intervals of 600, 900, 1,200, and 1,500 ms. We recorded the response times to GO trails, omissions  and intrusions ; (2) Stroop task. Participants were presented with letter strings for which they had to name the color. Each letter string could belong to either one of three conditions: congruent , incongruent , or neutral. This condition was represented by the letter string XXXX written in red, blue or green. Congruent and incongruent color words were VERDE, BLU, and ROSSO. There were 96 trials, with 24 congruent trials, 24 incongruent trails, and 48 neutral trials (16 for each color). We recorded the latency between presentation of the item and onset of the participant's response. The inter-trial interval was 2,000 ms. Prior to the experimental trials, there were 12 trails of practice in which participants named the color white in which VERDE, BLU, or ROSSO were written on the screen. For each participant, we computed a facilitation effect , an inhibition effect , and a maximal inhibition effect (incongruent RT minus congruent RT), (3) Intrusions errors in the Listening Span test (see below). They provide an estimate of inhibitory control since participants are required to use it to reduce the accessibility of non-final words, prevent their recall and therefore produce fewer intrusions  Forward Digit Span Test and Backward Digit Span Test  Phonemic and Semantic Fluency Tests  Spielberger State-Trait Anxiety Inventory  on a white background and remained on the screen until participant response. If no response was provided in 4,000 ms, the target disappeared and the next trial appeared. The participant response was followed by an interval of 900 ms before the subsequent sentence began. The experimental instructions were displayed on the screen and then repeated by the experimenter before the training session. Participants were informed that a letter string would appear at the end of each sentence. They had to decide whether it corresponded or not to an Italian word and press an appropriately labeled response key . After 10 training trials, the experiment began. Participants were instructed to pay attention to the sentences because they would be asked some questions at the end of the session. Immediately after the experiment, participants performed the recognition test. The experiment lasted on average 25 min.Testing and cross-modal lexical decision experiment were individually performed in two different sessions taking place a week after the other. The order of testing and experiment was quasi-randomized across participants. In one session, participants performed the following tests in this order: Listening Span test, Simple reaction times, Go-No Go task, Digit Span, Semantic and Phonetic fluency, Vocabulary; in another session, they carried out the cross-modal lexical decision experiment, the Stroop test, RPM, STAI-Y, and BFO.We conducted analyses of variance (ANOVAs) on the average reaction times of correct responses for condition to the cross-modal lexical decision experiment considering both participants and items as random factors. In the by-participant analysis (indicated by subscript 1), Idiom type (Predictable vs. Unpredictable) and Target (Idiomatic vs. Control) were within-subject factors. In the by-item analysis (indicated by subscript 2), Idiom Type was a between-subject factor and Target a within-subject factor.To explore whether individual difference in the cognitive and personality components were associated with the comprehension of idioms, we computed: (1) Pearson correlations between the participants scores in each test and their correct average response times to idiomatic and control targets in predictable and unpredictable idioms; to idiomatic and control targets regardless of idiom type; and to predictable and unpredictable idioms regardless of target type . Errors represented only the 1.07% (SD = 0.58) of the total responses, therefore they were not statistically analyzed . In order to reduce variability, data points ±2 analyses .3%. ErroF1 = 42.844, MSE 402590.25, ηp2 = 0.405, p = 0.000; F2 = 8.582, MSE 107976.446, ηp2 = 0.248, p = 0.007; MinF′  = 7.336, p = 0.01] with faster lexical decision times for idiomatic than for control targets (95% CI: 65± ms). Unfortunately, both the Idiom type factor and the Idiom Type X Target interaction were significant in the by-participant but not in the by-item analysis . The relatively low number of stimuli per condition is presumably responsible for the lack of by-item significant effects in both cases.The Target factor was significant . Idiomatic and control targets were both responded to faster in predictable than in unpredictable idioms .Bonferroni-corrected In sum, we found contextually-driven activation of the idiomatic meaning for both predictable and unpredictable idioms in line with our prediction and with previous studies . We only report the variables that led to statistically significant correlations with the measures of idiom processing Table . For sakr = −0.288, p = 0.021, r = −0.343, p = 0.005, respectively) and unpredictable idioms . In sum, the faster participants were in correctly responding to idioms, the higher was their verbal WM. The number of intrusions is positively correlated with the response times to idiomatic targets of predictable idioms  in that the faster participants were in responding to idiomatic targets, the lesser intrusion errors they made.Significant correlations only concerned the Listening Span test. Specifically, the number of recalled words inversely correlates with the response times to idiomatic and control targets of predictable (p = 0.012), to predictable and unpredictable idioms , regardless of target type, and with the response times to idiomatic and control targets, regardless of idiom type . In sum, the better the participants performance was in the Vocabulary test, the faster participants were in responding to idioms.Vocabulary scores inversely correlate with the response times to idiomatic targets of predictable idioms , with the response times to unpredictable idioms, regardless of target type , and with the response times to idiomatic and control targets, regardless of idiom type . In sum, higher levels of temporary anxiety due to a specific event (presumably the experiment) were associated to faster response times to these targets. Trait Anxiety scores inversely correlate only with the response times to the control targets of unpredictable idioms . Extraversion scores of the BFO inversely correlate with the response times to control targets of unpredictable idioms . Both Agreeableness and Openness to Experience scores inversely correlate with the response times to idiomatic and control targets of unpredictable idioms , with the response times to unpredictable idioms, no matter the target type , and with the response times to idiomatic targets, no matter the idiom type . Openness to Experience also inversely correlates with the response times to idiomatic targets of predictable idioms . In sum, the higher were the scores of Agreeableness and Openness to Experience, the faster were the response times to unpredictable idioms, and in general to control targets; the higher the scores of Openness to Experience, the faster the responses to idiomatic targets of predictable idioms, and lastly the higher the scores of Extraversion, the faster the responses to control targets of unpredictable idioms.State Anxiety scores inversely correlate with the response times to both idiomatic and control targets of unpredictable idioms  = 4.603, p = 0.001, R2 = 0.291] with significant contributions of the constant term , and of the correctly recalled words and intrusions of the Listening Span . Two further predictors had close-to-significance values: Openness to Experience and State anxiety . Block 3  accounted for an additional 6.6% of variance . The tolerance values  and the variance inflation factor values  did not suggest any multicollinearity in the analysis.In the hierarchical regression analyses, significant effects were observed for the response times to idiomatic targets of predictable idioms. Block 2  accounted for 29.1% of the variance , though the complete model was not significant , and to idiomatic targets of unpredictable idioms, where Vocabulary accounted for 8.1% of the variance , though again the complete model was not significant . None of the Blocks or single predictors produced significant R2 changes in the response times to the other experimental conditions.Significant effects of a single predictor (Vocabulary) were found in the response times to control targets of predictable idioms, where Vocabulary accounted for 10.1% of the variance [In line with prior evidence and with our predictions, when ambiguous idioms were embedded in idiom-biasing contexts the idiomatic meaning was equally available for predictable and unpredictable idioms. Clearly, the joint effects of the contextual bias and familiarity facilitated semantic processing making the idiomatic meaning available at the end of the idiom strings regardless of predictability.The correlation and regression analyses showed important contributions of WM, inhibitory control and crystallized verbal intelligence, as well as marginally significant contributions of State Anxiety and Openness to Experience. These results represent an interesting, although still exploratory, addition to the literature and extend to idioms previous evidence about the components associated with metaphor and sarcasm comprehension Blasko, . SpecifiVocabulary scores contributed to the response times to predictable and unpredictable idioms. The Vocabulary subtest of WAIS-R provides an estimate of global verbal intelligence function . Presumably, this (unpleasant) temporary emotional arousal led participants to recruit more cognitive resources speeding up idiom processing  and participants testing specific hypothesis about each component of the cognitive and personality architecture underlying language processing are required before we can ultimately decide whether the complex architecture subserving idiom processing totally or partially coincides with that used to comprehend literal language, and what are the roles and weights of each component in idiom comprehension. One might also wonder whether these results are specific to idioms or they are generalizable to other kinds of literal and non-literal word strings. Again this is an empirical question that requires future studies to be properly answered.Since this study used ambiguous idioms, the disambiguation of the idiom meaning required the presence and use of contextual information to decide which of the two meanings of the string was appropriate. Of course providing idiom-consistent information has facilitated idiomatic meaning activation speeding up response times  and non-compositional multi-word strings  (e.g., Arnon and Snider, Why it is important to study the role of individual differences in language acquisition and processing not relegating them to error variance (Kidd et al., CC initially conceived the idea for the study which was then further developed and finalized by CC and PC. The stimulus materials were prepared by PC. Data collection was made by PC. Analyses were run by CC, FF, and PC. The majority of this paper was written by CC.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Dear Sir,A 4-year old boy presented with pain abdomen, bilious vomiting and constipation for three days. On examination, abdomen was distended and tender. X-ray abdomen revealed multiple air-fluid levels suggestive of intestinal obstruction. Ultrasonography abdomen revealed dilated bowel loops with minimal free fluid in peritoneal cavity. On exploration, there was a band causing obstruction of terminal ileum which was released. Band was arising from Meckel’s diverticulum which was connected to a cyst at umbilicus. On further exploration, the cyst was connected to urachus which was patent . There was no communication between bladder and urachus, cyst and urachus, Meckel's diverticulum and the cyst. Resection of Meckel’s diverticulum with ileum and end to end ileo-ileal anastomosis was done. Urachus and cyst were also excised. Patient recovered well postoperatively. Histopathology confirmed urachal cyst, urachus, and Meckel’s diverticulum. Umbilicus is common passage for various structure during intrauterine life. These structures usually involute and their persistence result in a variety of anomalies at this site. Persistence of vitello-intestinal duct is common anomaly.[1] Urachal anomalies usually present as urachal sinus, patent urachus, or urachal cyst. Various authors have reported concurrent presence of both vitello-intestinal duct and urachal anomalies presenting with various features including intestinal obstruction and persistent discharge from the umbilicus.[2-5] Our case was unique as we found a Meckel’s diverticulum and a band which caused intestinal obstruction. Moreover, there was a urachal cyst at umbilicus which was attached to Meckel’s diverticulum on one side and to urinary bladder on other side through a patent tract which was not communicating with the urinary bladder. This was quite unusual. Source of Support: NilConflict of Interest: None declared


Volume of interest template was applied to measure specific binding ratios (SBRs) of caudate nucleus, putamen, and striatum representing DAT availability as follows; SBR = (target– cerebellum)/cerebellum. Plasma EGF was negatively correlated with the availabilities of both caudate nucleus , and putamen . After dividing subjects according to Apo E genotyping, DAT availability of caudate nucleus of Apo e4 non-carriers (n = 60) showed the positive correlation with cerebrospinal fluid (CSF) α-synuclein . Plasma EGF was negatively correlated with DAT availabilities of Apo e4 non-carriers. Further studies are needed to clarify underlying mechanisms of this phenomenon.We aimed to evaluate the association between plasma epidermal growth factor (EGF) and the availability of dopamine transporter (DAT) measured from  EGF is known to involve in the development of the nervous system, stimulating proliferation, migration, differentiation of neuronal cells, enhancing survival, and inhibiting apoptosis2. Supplement of EGF to Parkinson disease (PD) model rat prevented the dopaminergic neurodegeneration3. Decreased level of EGF was found in striatum of patients with PD as compared with controls in postmortem study3. Several studies reported that low plasma EGF was correlated with cognitive decline in PD patients and the high conversion rate to Alzheimer’s disease (AD)5.Epidermal growth factor (EGF), 6 kDa protein made up of 53 amino acids, is found at high concentrations in bile, urine, milk, and prostate fluid, at medium concentrations in tears, follicular fluid, sperm, and seminal plasma, and at low concentrations in plasma, serum, and saliva7. The loss of dopaminergic neuron is parallel to the level of expression of the dopamine transporter (DAT) mRNA6. DAT is on the presynaptic dopaminergic nerve terminal and controls dopamine levels by active reuptake of dopamine from the synaptic cleft8. As 123I-FP-CIT reflects the striatal DAT density8, the availability of DAT measured from 123I-FP-CIT single-photon emission computed tomography (SPECT) can be used in evaluating the neurodegenerative disease9.PD is a clinical syndrome showing bradykinesia, tremor, rigidity, and postural instability. It is characterized by the loss of dopaminergic neuron of the substantia nigra, and the presence of intraneuronal cytoplasmic inclusion123I-FP-CIT SPECT in healthy controls in this study.Although the effect of EGF in neurodegenerative disease is well documented in previous reports, the correlation of EGF with DAT in healthy controls has not been investigated yet. Therefore, we evaluated the association between plasma EGF and the availability of DAT measured from www.ppmi-info.org/data). For up-to-date information on the study, visit www.ppmi-info.org10. The study population consisted of 192 healthy controls with screening 123I-FP-CIT SPECT. According to PPMI criteria of healthy subjects, males or females with their age of 30 years or older at screening was included, and subjects with a neurological disorder, a first degree relative with idiopathic PD, Montreal Cognitive Assessment score of 26 or less, medications that might interfere with DAT SPECT scans, anticoagulants that might preclude safe completion of the lumbar puncture, or investigational drugs, and a condition that precludes the safe performance of routine lumbar puncture were excluded. Subjects without CSF biomarkers, plasma EGF, and Apo E genotyping were excluded. Medical history, cerebrospinal fluid (CSF), plasma EGF, and 123I-FP-CIT SPECT scans were downloaded. The PPMI study was approved by the local Institutional Review Boards of all participating sites  and written informed consent was obtained from each subject at the time of enrollment for imaging data and clinical questionnaires. All methods were performed in accordance with the relevant guidelines and regulations.Data used in the preparation of this article were obtained from PPMI database , and total tau were analyzed with multiplex xMAP Luminex platform , and Innogenetics immunoassay kits . α-synuclein were analyzed using an enzyme-linked immunosorbent assay . Plasma levels of EGF were measured by enzyme-linked immunosorbent assay  according to manufacturer instructions. Samples were run in duplicate and data used for this study met quality control measures for technical performance.CSF biomarkers of AβApo E genotyping was performed on DNA samples. Two non-synonymous single nucleotide polymorphisms, rs429358, and rs7412 were genotyped in each sample to distinguish between Apo e2, e3, and e4 alleles using TaqMan assays .123I-FP-CIT SPECT was performed during the screening visit for all subjects. SPECT scans were acquired 4 ± 0.5 hrs after injection of 111–185 MBq of 123I-FP-CIT. Subjects were pretreated with iodine solution or perchlorate prior to injection to block thyroid uptake. Raw data were acquired into a 128 × 128 matrix stepping each 3 or 4 degrees for the total projections. Raw projection data were reconstructed using iterative ordered subset expectation maximization with HERMES . The reconstructed images were transferred to pmod  for subsequent processing including attenuation correction.123I-FP-CIT template11. Specific binding of 123I-FP-CIT regarding DAT was calculated using a region of interest analysis. A standard set of volume of interest (VOI) defining caudate nucleus, putamen, and striatum (caudate nucleus + putamen) based on the Automated Anatomical Labeling (AAL) atlas12. The cerebellum was chosen as a reference region. VOI template was applied to measure specific binding ratios (SBRs) of caudate nucleus, putamen, and striatum representing DAT availability as follows; SBR = (target– cerebellum)/cerebellum.Downloaded scans were loaded using pmod v3.6  with Normality was examined using D’Agostino-Pearson omnibus test. Spearman correlation was used to measure the relationship of SBRs with CSF biomarkers, and plasma EGF. Mann-Whitney test was applied to compare SBRs, CSF biomarkers, and plasma EGF between Apo e4 non-carriers and carriers. Statistical analyses were performed using GraphPad Prism 7 for Mac OS X .2. Twenty-one subjects were Apo e4 carriers (25.9%). DAT availabilities in caudate nucleus  showed a reduction with aging as expected. When subjects were divided according to Apo e4 genotyping, Aβ1–42 was higher in Apo e4 non-carriers than Apo e4 carriers. However, age, sex, BMI, α-synuclein, p-Tau181, total tau, and plasma EGF showed no significant differences between Apo e4 non-carriers and Apo e4 carriers. Subjects’ characteristics are summarized in Table 81 healthy subjects  were included in this study. Mean age was 62.3 years. Mean BMI was 26.5 kg/mNone of CSF biomarkers showed significant association with DAT availability Table . Howeverwww.ppmi-info.org/data).Data used in the preparation of this article were obtained from PPMI database  to AD5. Jiang et al. demonstrated that plasma EGF was decreased in the early stage of PD and there was no significant difference of plasma EGF between advanced PD and normal control15. In study regarding EGF and dopamine uptake in cultured rat astrocytes, they suggested the existence of Na + -dependent and Na + -independent dopamine uptake in cultured rat astrocytes and they concluded that EGF might stimulate the expression and translocation of the extraneuronal DAT16. In study assessing EGF-ErbB1 action on developing midbrain dopaminergic neuron, authors showed that EGF elevated DAT level in mesencephalic cultures and they reported that EGF receptor inhibitor reduced DAT level in the striatum, nucleus accumbens, and globus pallidum in neonatal rats17. EGF levels in both striatum and serum of patients with chronic schizophrenia were reduced comparing with those of normal controls in postmortem study18. Laakso et al. showed that striatal DAT availability measured by18F-CFT was reduced in patients with chronic schizophrenia as compared with normal controls19. Thus, we expected that the DAT availability would be decreased if the level of plasma EGF was decreased. Unlike our expectations, negative correlation between plasma EGF and DAT availability was observed in healthy subjects. There was no relevant study to explain this negative correlation.EGF is known to have protective effect in dopaminergic neuron21, which has a function in suppression of apoptosis, regulation of glucose levels, modulation of calmodulin activity, chaperone activity, and regulation of dopamine biosynthesis20. Also, it has been known to be associated with neurodegenerative disease such as PD, dementia with Lewy bodies, MSA, and AD21. Especially in PD, α-synuclein interacts with tubulin, parkin, dopamine receptor, synphilin-1, phospholipase, and small ubiquitin related modifiers20. Previous studies showed that α-synuclein was not different between Apo e4 carriers and Apo e4 non-carriers in normal subjects, MCI, and AD, consistent with this study23. However, molecular linkage between Apo E and α-synuclein was demonstrated in one study using A30P and A53T transgenic mice. α-synuclein induces neuronal degeneration leading to Apo E deposition in spinal cords, astrocytes, and activated microglia of transgenic mice24. Astrocyte-secreted Apo E reduced α-synuclein uptake, and the effect was seen in Apo e4, followed by e3, and e225. Wersinger et al. reported that α-synuclein reduced DAT activity by recruitment of DAT from plasma membrane to cytoplasm in their in vitro study26. The decreased activity of DAT was caused by reduced dopamine uptake velocity, not by decreased DAT expression27. Kovacs et al. showed that the density of DAT, identified by immunochemistry, inversely correlated with the density of α-synuclein in the substantia nigra of patients with Lewy body disease and PD28. However, Bellucci et al. demonstrated that transgenic mice producing human α-synuclein had increased levels of striatal DAT29.α-synuclein, 14 kDa protein, consists of three domains with N-terminal lipid-binding α-helix, amyloid-binding central domain, and C-terminal acidic tail30. Among them, Apo e4 is known to be the greatest risk factor for AD, followed by Apo e3, contrary to protective effect of Apo e231. Apo E has a major role in metabolism of Aβ, which is abundant in brain of Apo e4 carriers than Apo e4 non-carriers31. In this study, CSF Aβ1–42 was higher in Apo e4 non-carriers than Apo e4 carriers. Consistent with this study, Prince et al. showed decreased amount of CSF Aβ in Apo e4 carriers in both AD and normal control groups32. α-synuclein attenuated the effect of EGF by showing decreased luciferase activity in a study by Iwata et al.33. Therefore, Apo e4 non-carriers may have higher CSF Aβ1–42, CSF α-synuclein, and CSF α-synuclein leading to both lower plasma EGF and higher DAT availability.Apo E has three isoforms which are known to Apo e2, Apo e3, and Apo e41–42, α-synuclein, plasma EGF, and DAT availability in healthy controls. However there are several limitations in this study. First, subjects included in this study was collected from PPMI database. Second, the number of subjects was small in Apo e4 carriers. Third, as PPMI database was collected from multiple institutes, the difference in image acquisition may affect the results.This is the first study that investigated the association between Apo E genotyping, AβIn conclusion, plasma EGF was negatively correlated with DAT availabilities of Apo e4 non-carriers. DAT availability was positively correlated with α-synuclein in Apo e4 non-carriers. Further studies are needed to clarify underlying mechanisms of this phenomenon.


Birth injuries are a diverse set of traumas afflicting a newborn during labor and/or delivery. These range from temporary paralysis to hematomas. Herein, a comprehensive review of the birth injuries is presented, including the risk factors, classification of various paralyzes and nerve damage, as well as bleeding complications. The predicted outcomes and complications, as well as the treatment options for various birth injuries, are also discussed. A birth injury is defined as structural damage or functional deterioration of a newborn secondary to a traumatic event that occurred during labor, delivery or both. Amniocentesis and intrauterine transfusions can cause injuries before birth, and therefore are not considered as birth injuries. The injuries following neonatal resuscitation procedures are also not classified as birth injuries. Some injuries can take place from use of fetal scalp electrodes and intrapartum heart rate monitoring and are characterized as birth injuries . The neoRisk factors for birth traumaThere are many identifiable risk factors that increase the likelihood of neonatal birth injuries. These can be classified into three broad categories: fetal presentation, delivery mechanisms, and maternal factors. The fetal factors include macrosomia, breech presentation, abnormal fetal presentation, prematurity and precipitous delivery. Macrosomia results in a number of related injuries such as shoulder dystocia, rib fractures, clavicular fractures, cephalohematoma, and caput succedaneum. The poorly controlled maternal diabetes is one of the leading causes of macrosomia . A birthDelivery mechanisms include obstetric instrumental techniques such as forceps delivery, vacuum extraction, or a combination of both, which increases the ease of descent but can potentially induce injury . The vacMaternal factors include age, number of pregnancies, and pelvic anatomy. Extreme maternal ages (less than 16 and greater than 35 years), primigravida, cephalopelvic disproportion, short maternal stature, and maternal pelvic abnormalities predispose a fetus to birth injuries [Obstetric brachial plexus injuriesObstetric brachial plexus palsy is defined as a flaccid paresis of an arm at birth, with the passive range of motion being greater than the active range of motion . The incThe patients with permanent obstetric brachial plexus injuries often develop bony asymmetry as a result of muscle imbalance on the developing bony elements of the infant’s shoulder . The exaThe birth weight among the injured patients tend to be higher than average, but it is not associated with injury severity. Nath, et al.  found thBrachial plexus injuriesThe classic obstetric brachial plexus birth injury is Erb’s paralysis and is more common than Klumpke’s palsy and facial nerve injury. The injury can consist of two or more spinal roots. The degree of severity can vary from neuropraxia and axonotmesis to more serious injuries such as root avulsion . The excThe treatment is aimed at preventing contractures and can range from splinting and encouraging passive exercises in mild cases to the surgical exploration and grafting in more serious cases or for those who do not show recovery by three months , 20-21. Erb’s paralysisErb’s paralysis or Erb’s palsy occurs as a result of damage to the C5-C6 motor roots, in which the C5 and C6 unite to form the upper trunk of the brachial plexus . The bicWhen damage to C8 and T1 are included, there is total limb paralysis . This isKlumpke’s paralysisKlumpke’s paralysis involves C7, C8 and T1 nerve roots which results in the loss of function of the intrinsic muscles of the hand as well as the wrist and finger flexors -24. The Facial nerve paralysisFacial nerve paralysis can result from perinatal trauma. The main reported risk factors associated with traumatic facial paralysis are primigravida, birth weight greater than 3500 g, forceps usage, cesarean birth, and prematurity. These cases typically have a good prognosis, with infants recovering without sequelae . The mecIntracranial hemorrhageTrauma during delivery is the main causes of intracranial hemorrhage. The risk factors include excessive molding of the head in prolonged labors, breech deliveries, and use of forceps . Full-teThe clinical presentation depends on the extent of the bleed. Accumulation of blood in the posterior fossa can cause a tense or bulging fontanelle, which increases the head circumference and predisposes to central sleep apnea, bradycardia, and/or seizures. Stillborn births are a severe complication . In caseHematomasHematoma of the Sternocleidomastoid MuscleThe use of forceps in breech deliveries can cause the rupture of the sternocleidomastoid muscle (SCM) fibers and blood vessels, leading to a hematoma and subsequent development of a fibrous mass appearing at two to three weeks of age, resulting in a slight torticollis , 31. TheCaput SuccedaneumCaput succedaneum is described as swelling that occurs due to the increased pressure of the vaginal and uterine walls on the fetal head during labor . This prCephalohematomaCephalohematoma is caused by a subperiosteal collection of blood due to rupture of vessels beneath the periosteum. The occurrence rate is 1-2% of all the deliveries regardless of the mode of delivery. Most cases are unilateral due to the fact that bleeding does not cross the suture lines as it occurs within a single cranial plate. Due to slow bleeding, the clinical presentation of swelling is delayed for several hours to days. A tense mass with ridge at the margin of the lesion and a recessed center is palpable. The lesion increases in size after birth and is usually largest on the third day of postnatal life . CephaloSubgaleal HemorrhageSubgaleal hemorrhage is a collection of blood between the epicranial aponeurosis and periosteum of the skull. The risk of the subgaleal hemorrhage is greater with vacuum extraction deliveries . The subIt is important for the clinicians to understand the range of injuries that can occur to a newborn during birth. The fetal presentation, delivery mechanisms, and maternal factors can each influence the risk of a neonatal birth injury as well as the kind and severity of such injury. Such risk factors should be taken into consideration by obstetricians when dealing with complications during birth. Understanding the risk factors as well as the various classifications of the birth injuries ensures the best possible outcome for the newborn.


Z score > 0.5) or decelerated (Δweight ΔZ < −0.5) growth during early (0–3 months) and late (3–12 months) postnatal life on metabolic health indicators at age 1-year. The primary outcomes were homeostasis model assessment of insulin resistance (HOMA-IR), β-cell function [homeostasis model assessment of β-cell function (HOMA-β)], and fasting plasma lipids. Adjusting for maternal, paternal, and infant characteristics, accelerated growth during the first 3 months of life was associated with a 41.6%  increase in HOMA-β, and a 8.3% (0.7–15.4%) decrease in fasting plasma total cholesterols, and was not associated with HOMA-IR in infants at age 1-year. Accelerated growth during 3–12 months was associated with a 30.9% (3.3–66.0%) increase in HOMA-IR and was not associated with HOMA-β. Neither accelerated nor decelerated growth was associated with fasting plasma triglycerides, high-density lipoprotein or low-density lipoprotein cholesterol concentrations in infants at age 1-year. Accelerated growth during early postnatal life may be beneficial for β-cell function, but during late postnatal life harmful for insulin sensitivity in normal birth weight infants.Accelerated growth in postnatal life in low birth weight infants has been associated with insulin resistance and metabolic syndrome-related disorders in later life. Postnatal accelerated growth in also common in normal birth weight infants, but little is known about the impact on metabolic health. In a prospective cohort study of 203 term normal birth weight infants, we evaluated the impacts of accelerated (Δweight  Z score > 0.5 or 0.67) which has been consistently associated with insulin resistance and metabolic syndrome-related disorders  in later life  of type 2 diabetes in adulthood , the vasn = 307), and the infants were followed up at 3 months (n = 280) and 1 year (n = 241) of age. This study included all 203 normal birth weight babies with postnatal growth data and fasting infant blood specimen available for biomarker assays  of age, infants were followed up for data collection on feeding and growth measurements. Infant feeding was classified as any breastfeeding or non-breastfeeding. Infant’s length was measured in supine position by the 447 Infantronic Digital Infantometer  to the nearest 0.1 cm. Weight was measured by an electronic weighting scale to the nearest gram (g). Skinfold thickness at triceps and subscapular positions was measured by a Harpenden skinfold caliper  to the nearest 0.1 mm. All anthropometric measurements were taken twice, and the average values were used in the final analysis data. All follow-ups and anthropometric measurements were done by a trained pediatric research nurse.z scores  at birth, 3, and 12 months of age were calculated using the WHO child growth standards  tube of blood was specifically collected for the glucose assay. The blood specimens were kept on ice, and centrifuged within 30 min after specimen collection. The separated plasma samples were stored in multiple aliquots in a −80°C freezer until assays.p > 0.1).The assays of plasma glucose and insulin followed the protocols described previously . Plasma z score [WHO child growth standards (The primary outcomes were homeostasis model assessment of insulin resistance (HOMA-IR) and β-cell function [homeostasis model assessment of β-cell function (HOMA-β)] and fasting plasma lipids  in infants at age 1-year. HOMA-IR was calculated as fasting insulin (mU/l)*fasting glucose (mmol/l)/22.5, HOMA-β as fasting insulin (mU/l)/[fasting glucose (mmol/l) −3.5] . Other otandards ], body mtandards .z score), and breast feeding (yes/no). Biomarker variables (positively skewed crude data distributions) were log-transformed in the comparisons for differences between groups. Generalized linear regression models were used to assess the differences adjusting for multiple co-variables. For log-transformed biomarker outcomes, we calculated the adjusted % differences in the original scale between two groups according to the regression coefficients. The sample size is sufficient in multivariate regression models since linear models require a minimal of two subjects per variable , pre-pregnancy BMI (SD score), biological father’s height (SD score) and BMI (SD score), maternal family history of diabetes , ethnicity , age , parity (primiparous: yes/no), prenatal smoking (yes/no), alcohol use (yes/no), gestational diabetes (yes/no), gestational hypertensive disorders (yes/no), mode of delivery , infant sex, gestational age (weeks), birth weight (variable . Data man = 203). The majority of mothers were white (about 70%). About a quarter of mothers were older than 35 years, and 17% of mothers had a family history of diabetes. The mean pre-pregnancy BMI was 23.8 kg/m2. The mean birth weight was 3,458 g. About 90% of infants were breast-fed. The average changes in weight z scores were close to 0 during both 0–3 and 3–12 months of postnatal age.Table Table p = 0.01), an 8.3% (0.7–15.4%) decrease in fasting plasma TC (p = 0.03), and a 2.7  mm increase in the sum of triceps and subscapular skinfold thickness, but was not associated with HOMA-IR in infants at age 1-year. Accelerated growth during 3–12 months of life was associated with a 30.9% (3.3–66.0%) increase in HOMA-IR (p = 0.03), and a 1.5  mm increase in the sum of triceps and subscapular skinfold thickness, but was not associated with HOMA-β. As expected, during both 0–3 and 3–12 months of age, accelerated growth was associated with higher BMI and weight-for-length z scores, while decelerated growth was associated with lower BMI and weight-for-length z scores in infants at age 1-year.Table Our study is the first to reveal potentially differential metabolic health impacts of accelerated growth during early vs. late postnatal periods during the first year of life in normal birth weight infants. Accelerated growth during the first 3 months appears to be beneficial for β-cell function, but during 3–12 months harmful for insulin sensitivity in infants at age 1-year.n = 203), we first discovered the timing-dependent impacts of accelerated growth in early (first 3 months) vs. late postnatal (3–12 months of age) periods on metabolic health in infancy. Accelerated growth during the first 3 months of life was associated with higher β-cell function and lower fasting blood TC, but during 3–12 months of life was not. Crude mean HOMA-β was 31.9% higher in infants with accelerated growth compared to those with normal growth during the first 3 months of life (mean: 118.3 vs. 86.4%). This difference did not reach statistically significance (p = 0.11) in the crude comparison (Table p = 0.01) after the adjustments  and 23 birth weight appropriate for gestational age (AGA) infants, and found that accelerated growth in SGA infants during the first year of life was associated with increased insulin resistance . In contAs expected, accelerated growth during either early or late postnatal period was associated with greater skinfold thickness at age 1-year. Accelerated weight gain may come along with some fat deposition in subcutaneous tissue.We are unaware of any research data on the relationship between decelerated growth and metabolic health in normal birth weight infants. As expected, decelerated growth in either the 0–3 or 3–12 months was associated with lower BMI at age 1-year. Decelerated growth during 3–12 months was also associated with lower skinfold thickness. However, decelerated growth in either period was not associated with HOMA-IR or HOMA-β.The observed positive association between early postnatal accelerated growth and HOMA-β in infant at age 1-year is consistent with the findings in animal studies. It has been shown that insulin secretion in 20-day-old mice was decreased significantly as a consequence of restricted nutrient intake during the suckling period , and greStrengths of the study include the high rates of infant follow-up and fasting blood specimen collection, and the analyses accounting for prenatal and postnatal potential confounding factors. One limitation is that the HOMA insulin resistance and β-cell function indices are based on glucose and insulin concentrations in a single fasting blood sample. Studies using more invasive techniques with multiple blood samplings  may yield more accurate estimates of insulin resistance and β-cell function, but such procedures are not so acceptable to most parents of small infants. However, HOMA-IR and HOMA-β have been validated against the gold standard methods—the euglycemic–hyperinsulinemic clamp and first-phase insulin secretion on intravenous glucose tolerance test in children . The stuAccelerated growth during early postnatal life (the first 3 months) may be beneficial for β-cell function, but during late postnatal life (3–12 months) detrimental for insulin sensitivity in normal birth weight infants at age 1-year. The findings suggest differential impacts of early vs. late postnatal accelerated growth on metabolic health in normal birth weight infants.This study complies with the guidelines of the Declaration of Helsinki. Written informed consent has been obtained from each study participant. The study was approved by the research ethics board of Sainte-Justine Hospital Research Center, University of Montreal.Z-CL, PJ, and WF developed the research protocol and obtained the research grants; D-LZ, QD, AD, PJ, WF, and Z-CL contributed to the acquisition of research data; D-LZ, QD, and Z-CL contributed to the data analyses; D-LZ drafted the manuscript. All authors contributed to improvements of the manuscript for important intellectual content and approved the final version for publication.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Three studies so far have investigated the effect of prenatal non-steroidal anti-inflammatory drug (NSAID) exposure on birth weight and gestational age. The aim in this study was to evaluate the association of prenatal ibuprofen with birth weight and gestational age at birth, using a sibling design in an attempt to adjust for the possibility of familial confounding.Using data from the Norwegian Mother and Child Cohort Study (MoBa) and the Medical Birth Registry of Norway (MBRN), we identified 28 597 siblings, of whom 1080 were prenatally exposed to ibuprofen and 26 824 were not exposed to any NSAID. Random and fixed effects models with propensity score adjustment were used to evaluate the effects of ibuprofen exposure on birth weight and gestational age.Ibuprofen exposure during the first trimester was associated with a decrease in birth weight of 79 grams . In contrast, second and/or third trimester exposure, and duration of exposure had no impact on the effect estimates. We found no association between ibuprofen exposure and gestational age at birth.Our results suggest that prenatal exposure to ibuprofen during the first trimester is associated with a slight decrease in birth weight. The association does not seem to be attributable to shared genetics and family environment, and could be explained by either exposure to ibuprofen, or to non-shared confounding between pregnancies. The prevalence of non-steroidal anti-inflammatory drug (NSAID) use during pregnancy ranges between 5% and 17%.–14, congSince thousands of pregnant women use NSAIDs worldwide, and most commonly ibuprofen, even a small effect on birth weight or gestational age could have a significant impact on public health. Although we in our previous study had suffThe aim of the present study was to reevaluate the association of prenatal ibuprofen exposure with birth weight and gestational age using a sibling design. Such a design allows for adjustment of genetic and familial confounding factors that are shared between siblings, because siblings share on average 50% of their genes and to a large degree share family environment. In orderThis population-based sibling study was based on The Norwegian Mother and Child Cohort Study (MoBa) and The Medical Birth Registry of Norway (MBRN). MoBa and MBRN were linked via the 11-digit maternal identification number assigned to every resident of Norway.MoBa is an ongoing observational prospective cohort study conducted by the Norwegian Institute of Public Health (NIPH). The prinAll pregnant women living in Norway who gave birth between 1999 and 2008 were invited to participate in MoBa. There were no exclusion criteria, and the participation rate was 40.6%., 31 InfoThe MBRN includes all pregnancy outcomes in Norway after the 12th gestational week. Information on maternal health both prior to and during pregnancy, the course of pregnancy and pregnancy complications, delivery and postpartum complications and interventions, and the health of the neonate is available from standardised mandatory forms completed by midwives and obstetricians and/or gynaecologists at each delivery and from antepartum obstetric records completed by general physicians, gynaecologists, or midwives throughout pregnancy.Of the initial population of 114 275 pregnant women, 32 946 participated more than once. After excluding multiple pregnancies, pregnancies lacking information in any of the two study questionnaires and siblings exposed to other NSAIDs than ibuprofen, a total of 27 904 siblings were included in the final study population. The final study sample included 1080 siblings (3.9%) exposed to ibuprofen during pregnancy; of these 996 belonged to a sibling pair and 84 were in clusters of three or more. The remaining 26 824 siblings (96.1%) were unexposed to any NSAID. Women included in our study sample had equivalent socio-demographic characteristics to those not included. A flow chart of the study population is shown in The mothers of the siblings included in the final study sample displayed similar socio-demographic characteristics compared with the mothers in the initial MoBa cohort participating with one child only. However, a statistically higher proportion of the mothers in the sibling study sample had a medical diagnosis before but not during pregnancy (56% in the sibling sample compared with 44% in the cohort).Information on the timing and duration of ibuprofen exposure was available from the two MoBa questionnaires answered during pregnancy (Qw17 and Qw30). The questionnaires were designed to increase recall and several indications where analgesics could be applied were specifically named: musculoskeletal pain (acute), inflammatory musculoskeletal conditions (both chronic and acute), infectious diseases (acute), neuromuscular inflammatory conditions (acute), and inflammatory conditions accompanied by symptoms of pain. An open-ended question about medications use was also included. For each indication, medication use in the following time periods could be specified: gestational weeks 0 to 4, 5 to 8, 9 to 12, and 13+ (Qw17) and weeks 13 to 16, 17 to 20, 21 to 24, 25 to 28, and 29+ (Qw30). Drug exposure was classified and grouped according to the Anatomical Therapeutic Chemical (ATC) Classification System developed by the World Health Organization. IbuprofeThe outcome variables in our study, birth weight in grams and gestational age in days were derived from the MBRN. Birth weights outside 3.5 standard deviations from the gender specific mean at each pregnancy week (0.5%) and gestational ages exceeding 44 weeks (0.9%) were recoded as missing. We tested the intra-class correlation for gestational age and birth weight to evaluate similarity within the sibling clusters.The following measured medically plausible confounding variables were included in the propensity score that was used as the adjusting variable: maternal age, parity, pre-pregnancy Body Mass Index (BMI), education, smoking throughout pregnancy, alcohol intake equal to or exceeding 1 unit per week during pregnancy, back pain, pelvic girdle pain, and neck and shoulder pain during pregnancy, migraine and headache during pregnancy, infections of the genitourinary tract during pregnancy, rheumatoid disorders, use of opioid analgesics during pregnancy, and birth order of the child. Some additional confounding variables were included for birth weight, only , and in addition, some were included for gestational age, only .To create the propensity score, we used logistic regression in which ibuprofen exposure was the dependent variable and the variables considered (i) material or theoretical confounders, or (ii) risk factors for the outcome, were predictors. The propensity score is the predicted exposure status, conditional on measured confounders; it has The analyses of the effect of ibuprofen exposure on birth weight and gestational age were carried out in three steps. First, a crude random effects analysis was performed to provide an estimate of the association between ibuprofen exposure during pregnancy and pregnancy outcome for the whole cohort of 28 597 siblings. This estimate is comparable to that obtained from a non-sibling design. Second, the propensity score variable was used as the adjusting variable in the random effects model. Third, a fixed effects model, adjusted for propensity score, was used to adjust for shared unmeasurable family-level effects as well as measured confounders included in the propensity score, see fixed effects formula figure . For allStatistical analyses were performed with Stata statistical software, version 14.Of the 27 904 siblings included in this study, 1080 (3.9%) were prenatally exposed to ibuprofen: 798 were exposed during the first trimester and 481 of these were not exposed to any other trimester; 249 were exposed during the second and/or third trimesters but not in the first, and 33 were exposed anytime during pregnancy without the particular trimester being specified. Regarding duration of use, there were 740 siblings exposed to ibuprofen during any one trimester and 340 siblings exposed to ibuprofen during two or more trimesters.The characteristics of the mothers who used ibuprofen during pregnancy and mothers who did not use any NSAIDs during pregnancy are presented in Tables The mean birth weights of infants exposed to ibuprofen anytime during pregnancy, during the first trimester only and during the second and/or third trimesters only were 3596 ± 585 grams, 3576 ± 573 grams, and 3660 ± 587 grams, respectively. The mean birth weight of infants born to women who did not use any NSAIDs during pregnancy was 3624 ± 567 grams. There were 31 infants (3.0%) exposed to ibuprofen during pregnancy that were born with a birth weight of less than 2500 grams, as compared to 558 infants (2.1%) who were not exposed to any NSAIDs during pregnancy.th gestational week, in contrast to 1034 infants (3.8%) who were not exposed to any NSAIDs during pregnancy.The mean gestational age (39.2 weeks) was the same for all infants whether they were exposed to ibuprofen anytime during pregnancy, during the first trimester only and during the second and/ or third trimesters or not exposed at all. There were 54 infants (5.0%) exposed to ibuprofen during pregnancy that were born before the 37The intra-class correlation within sibling clusters for gestational length and birth weight was 0.40 and 0.21 respectively, indicating a small effect of familial factors.Tables We did not find any association between ibuprofen exposure during pregnancy and gestational age.The primary finding in our study is that infants exposed to ibuprofen during the first trimester, including those born at term, had a birth weight that is on average 79 grams lower than their siblings not exposed to any NSAIDs during pregnancy.Our results are similar to those obtained in our previous cohort study where ibWe found no association between prenatal ibuprofen exposure and gestational age in the present study, thereby confirming our previous findings and those of the two other studies that assessed the effect of NSAIDs on gestational age., 13The birth weight of a neonate is likely to be influenced by factors occurring towards the end of pregnancy. However, recent studies are increasingly pointing to early pregnancy factors that particularly influence placental development, vascularization, and ultimately function. PregnancInterestingly, the observed effect estimate for birth weight is almost twice as large in our adjusted fixed effects model as the estimate obtained in the adjusted random effects model. There are several possible explanations for this difference. First, it is possible that unmeasured, shared familial confounders obscured the true effect of ibuprofen exposure, and the fixed effects model reduced the effect of these confounders. However, previous research has suggested the sibling studies may produce biased estimates in the case of non-shared confounding between pregnancies. Given thOur study has several strengths and weaknesses that merit attention. One strength is that the sibling design enables the adjustment of shared genetic and familial factors across pregnancies–something we were unable to do in our previous cohort study. In additThe limitations of our study also need to be taken into account. The low participation rate of 40% in MoBa is often a source of concern with respect to possible selection bias, especially with regard to prevalence estimates, as women younger than 25 years, without a life-partner, with a parity >0, and smokers are all under-represented, whereas mothers who take folic acid are over-represented. On the other hand, the fact that only minor differences (below 2% in absolute differences in socio-demographic variables) have been reported between MoBa participants and the general Norwegian population of pregnant women strengthens the representativeness of our material., 31 SincOur results suggest that prenatal exposure to ibuprofen during the first trimester was associated with a slight decrease in birth weight. The association does not seem to be attributable to shared genetics and family environment and could be explained by either the exposure to ibuprofen or changes in inter-pregnancy factors not accounted for in the analysis. No associations between ibuprofen exposure during pregnancy and gestational age were detected.The study was approved by the Regional Committee for Ethics in Medical Research, Region South, and the Norwegian Data Inspectorate.S1 Fig(TIF)Click here for additional data file.S1 TableThe β obtained in the random effects linear regression is representative of the β that would be obtained in a linear regression performed on the whole cohort. The fixed effects linear regression model addresses unmeasured and residual family-level confounding.* Propensity score variable for ibuprofen exposure during pregnancy included in all models .(DOCX)Click here for additional data file.S2 TableThe β obtained in the random effects linear regression is representative of the β that would be obtained in a linear regression performed on the whole cohort. The fixed effects linear regression model addresses unmeasured and residual family-level confounding.* Propensity score variable for ibuprofen exposure during pregnancy included in all models .(DOCX)Click here for additional data file.


To report a rare case of large radicular cyst-associated deciduous tooth and to discuss the importance of cone beam computed tomography (CBCT) in diagnosing the condition.Radicular cyst is the most common cyst affecting the permanent teeth, but its occurrence in deciduous teeth is rare. Most of the radicular cysts are asymptomatic and are discovered accidentally when radiographs are taken. Conventional radiographs show two-dimensional images of three-dimensional objects. Cone beam computed tomography provides undistorted three-dimensional information of hard tissues and gives adequate spatial resolution.A 7-year-old child, with a complaint of swelling in the maxillary anterior region, was diagnosed with radicular cyst in relation to primary maxillary right central incisor based on CBCT and histopathological features.Early diagnosis and prompt treatment of radicular cyst in primary dentition is important to prevent damage to permanent tooth.Mahesh BS, Shastry SP, Murthy PS, Jyotsna TR. Role of Cone Beam Computed Tomography in Evaluation of Radicular Cyst mimicking Dentigerous Cyst in a 7-year-old Child: A Case Report and Literature Review. Int J Clin Pediatr Dent 2017;10(2):213-216. These cysts usually involve the apex of the teeth.2 Most of the radicular cysts are asymptomatic and are discovered accidentally when periapical radiographs are taken. Radiographically, radicular cyst appears as a well-defined round or oval unilocular radiolucency with radiopaque sclerotic margin in the periapical region of involved tooth, but in case of an infected cyst, the radiopaque margin disappears because of rapid growth of the cyst.Radicular cyst is the most common cyst of the jaws, with frequency of about 7 to 54% in permanent dentition. But radicular cyst of deciduous dentition is extremely rare (0.5 to 3.3%).3 Cone beam computed tomography (CBCT) provides three-dimensional images of the object from sagittal, coronal, and axial direction, to overcome the defects of two-dimensional image, such as overlap and deformation.4 Hence, the aim of this study is to report a rare case of a large radicular cyst associated with maxillary right primary central incisor and to discuss the importance of CBCT in diagnosing the condition.The conventional radiographs show two-dimensional image of the three-dimensional object and its surrounding structures.A 7-year-old male patient reported to the Department of Oral Medicine and Radiology with the chief complaint of swelling in the upper front tooth region since 25 days. Swelling was small initially, which gradually increased to the present size. Swelling was associated with pain, which was continuous, dull, and mild in nature. Patient gave history of trauma to the upper front tooth region 2 years back. There was no history of pus discharge and ulceration associated with the swelling. On examination, extraorally diffuse swelling in the right maxillary anterior region in the middle third of the face was noted with upward displacement of nares . The swePatient was then subjected for radiographic investigation. Intraoral periapical (IOPA) radiograph and occlusal radiograph revealed proximal caries involving enamel and dentin without pulpal involvement in relation to 51, well-defined radiolucency in the periapical regions of 51 and 52, and this radiolucency was enveloping the coronal portion of 11 and 12. Root resorption of 51 and 52 were noted. Occlusal radiograph also revealed the presence of buccal cortical plate expansion .Cone beam computed tomography revealed osteolytic lesion with sclerotic margin in the periapical region of 51, measuring about 1.8 × 2.5 × 2.5 cm in dimension. The tooth 11 was displaced apically, root resorption of 51 and 52 was seen. Thinning of nasal floor with buccal cortical plate expansion was present  and 5. B5 Commonly, cyst is associated with carious, nonvital, discolored, or fractured tooth or teeth. However, in primary teeth, Grundy, Adkins, and Savage reported that teeth with endodontic treatment are more prone for development of radicular cyst. It was proposed that antigen stimulation by various products used for the endodontic treatment might be associated with development of a radicular cyst in primary teeth.6 Trauma can also be less commonly attributed etiology of development of radicular cyst in primary teeth.5The incidence of radicular cyst in deciduous tooth seems to be low as compared with permanent tooth. The probable reasons could be the shorter lifespan of primary teeth, and presence of numerous accessory canals leading to easy drainage in the primary teeth. Further, extraction or exfoliation of the associated deciduous tooth often leads to resolution of lesion on its own.7 Present case was reported in a 7-year-old male patient, involving maxillary primary incisor following trauma to the upper anterior teeth.Radicular cysts of deciduous teeth occur in the age group of 3 to 19 years, with male predominance. Most commonly reported teeth are mandibular molars (67%), maxillary molars (17%) followed by anterior teeth.8 Radicular cysts may also cause resorption, delayed eruption, malposition enamel defects, or damage to the developing permanent teeth.9Radicular cysts associated with primary teeth may cause expansion of cortical plates, especially buccal cortical plate.2 reported that the characteristic radiographic feature of dentigerous cyst is that pericoronal radiolucency is attached to cementoenamel junction of the associated teeth. Cone beam computed tomography clearly revealed that radiolucency was not attached at the cementoenamel junction of central incisor, but enveloping the periapical region of deciduous teeth. Thus, preopera-tive diagnosis of radicular cyst was substantiated. Further, during surgical procedure, the cyst was associated with deciduous tooth and permanent central incisor was pushed apically. Moreover, on histopathological examination, diagnosis of radicular cyst was confirmed.Radiographic changes of radicular cyst include well-defined radiolucency, expansion of cortical plates with thin reactive cortex, and displacement of permanent tooth. In the present case, IOPA and occlusal radiograph showed well-defined radiolucency in the periapical regions of 51 and 52, and this radiolucency was enveloping coronal portion of 11. But distinguishing whether this radiolucency was associated with 51 or 11 was difficult to determine. Cone beam computed tomography gave us the clear evidence of well-defined radiolucency, in the periapical region of 51, while the tooth 11 was displaced apically. This radiolucency was not attached to the cervical aspect of 11. Shear and SpeightDistinguishing between radicular cyst and dentiger-ous cyst is very essential as the treatment plan for both would be different. Cone beam computed tomography gives adequate three-dimensional views of oral and maxillofacial structures. In the present case, CBCT assisted in the preoperative diagnosis of radicular cyst, thus helping in the surgical planning.10 reported growth of 4 mm each year for radicular cyst, but in our case growth rate was faster. Treatment of radicular cyst in deciduous teeth involves complete enucleation of the cyst and preservation of the permanent successor teeth. Marsupialization is a conservative intervention and is done for larger cysts. But, with this technique, pathologic tissue is left in situ and multiple visits are required for regular draining of the cavity.11 In our case, extraction of deciduous teeth, cyst enucleation, and preservation of permanent teeth were done.HillConsidering the low incidence of radicular cysts in primary dentition, this case report informs possibility of a large radicular cyst secondary to trauma in maxillary anterior deciduous teeth and highlights the significance of CBCT in aiding the diagnosis of the condition because of adequate spatial resolution and undistorted hard tissue information. Early diagnosis and prompt treatment of radicular cyst in primary dentition is important to prevent damage to permanent tooth as well as the need for prevention of invasive surgical treatment.


Biological modelling of routinely collected newborn screening data has emerged as a novel method for deriving postnatal gestational age estimates. Validation of published models has previously been limited to cohorts largely consisting of infants of white Caucasian ethnicity. In this study, we sought to determine the validity of a published gestational age estimation algorithm among recent immigrants to Canada, where maternal landed immigrant status was used as a surrogate measure of infant ethnicity.We conducted a retrospective validation study in infants born in Ontario between April 2009 and September 2011.Provincial data from Ontario, Canada were obtained from the Institute for Clinical Evaluative Sciences.The dataset included 230 034 infants born to non-landed immigrants and 70 098 infants born to immigrant mothers. The five most common countries of maternal origin were India (n=10 038), China (n=7468), Pakistan (n=5824), The Philippines (n=5441) and Vietnam (n=1408). Maternal country of origin was obtained from Citizenship and Immigration Canada’s Landed Immigrant Database.Performance of a postnatal gestational age algorithm was evaluated across non-immigrant and immigrant populations.Root mean squared error (RMSE) of 1.05 weeks was observed for infants born to non-immigrant mothers, whereas RMSE ranged from 0.98 to 1.15 weeks among infants born to immigrant mothers. Area under the receiver operating characteristic curve for distinguishing term versus preterm infants  was 0.958 and 0.986, respectively, in the non-immigrant subgroup and ranged from 0.927 to 0.964 and 0.966 to 0.99 in the immigrant subgroups.Algorithms for postnatal determination of gestational age may be further refined by development and validation of region or ethnicity-specific models. However, our results provide reassurance that an algorithm developed from Ontario-born infant cohorts performs well across a range of ethnicities and maternal countries of origin without modification. This validation study has successfully demonstrated that a gestational age estimation algorithm performs well across infants from diverse backgrounds without modification.Population-based design: The validation cohort included 300 132 live-born infants born between April 2009 and September 2011 who underwent newborn screening at Newborn Screening Ontario.Defining ethnic subpopulations: Landed immigrant status, rather than self-reported ethnicity, was used as a surrogate measure for identification of ethnic subpopulations.Model development: The model demonstrates poorer performance among more severely preterm infants, in part due to smaller numbers of preterm infants available for model development.Knowledge of gestational age at the time of birth is vital for ensuring adequate provision of newborn care and for assessing population-level estimates of the burden of preterm birth to guide allocation of health services resources and targeted global health initiatives.in utero environmental exposure,Secondary analysis of newborn screening samples routinely collected within the first few days of an infant’s life has emerged as a unique opportunity for postnatal gestational age assessment. We and others have recently demonstrated the accuracy of postnatal gestational age algorithms derived from newborn screening data in three independent North American cohorts.We conducted a population-based retrospective validation study of infants born in Ontario, Canada using a combination of linked health administrative, newborn screening and immigration datasets maintained by the Institute for Clinical Evaluative Sciences. The Newborn Screening Ontario (NSO) database includes the analyte profiles of each infant completing newborn screening in the province . Over 40 screening analytes and analyte ratios including acylcarnitines, amino acids, endocrine markers, enzymes and coenzyme markers among others are available in the NSO database . MaternaThe postnatal gestational age estimation model previously published by our group included 249 700 infants born between April 2007 and March 2009. The details of the original model have been described previously.The data were randomly partitioned into model development (50%), validation (25%) and test (25%) subsamples. Multiple linear regression was performed in the model development set, in which all analyte main effects were included in the model, as well as birth weight and sex. For analytes and birth weight, squared and cubic terms were also included. A stepwise variable selection algorithm was then conducted and all pairwise interactions were considered for inclusion. The Schwartz Bayesian Criterion, which rewards improved model fit and penalises model complexity, was used in variable selection. Once this process was complete, the mean square error (MSE) for the fitted model at each step was calculated in the independent validation sample subset, and the model with the smallest MSE was selected. This approach provides a high level of protection from overfitting in the final model.The deviation of each calculated gestational age from the true gestational age of each infant is the residual model error for that infant . The residual model error can be positive or negative depending on the direction of the difference. The MSE is the mean of each of those residual errors after squaring it  for all infants. MSE is presented in the units of weeks.Gestational age was then dichotomised to distinguish term or preterm infant subgroups. Model performance to correctly classify infants across two thresholds of preterm birth categories by logistic regression analysis was assessed (area under the receiver operating characteristic curve (AUC), sensitivity, specificity and positive predictive value). Preterm birth thresholds were <37 weeks gestational age, the clinical threshold for preterm birth, and ≤34 weeks gestational age, a threshold that represents the lower limit of the late preterm period. Infants born ≤34 weeks gestational age (early preterm and severe preterm infants) are at increased health risk compared with those born late preterm or at term. Estimated gestational age identified by multiple linear regression was used as a continuous independent variable in logistic regressions to determine the probability of preterm birth.The validation cohort included 300 132 live-born infants born between April 2009 and September 2011 who underwent newborn screening at NSO. A total of 348 098 records were available for the prescribed study period. Records with missing or implausible (data entry error) data on birth weight, sex and gestational age were removed from the analysis. Infants who received a ‘positive’ flag for any one of the newborn screening conditions were excluded. Also excluded were those with missing newborn screening analyte values, or those for which the samples were collected before 24 hours and after 7 days of birth. The coefficients from the reference linear regression model were fixed, and used to score the validation cohort. Calculated gestational age was used as the independent variable in logistic regression models to estimate the dichotomous categories of preterm birth, for which model performance characteristics including AUC, sensitivity, specificity and positive predictive value were calculated. Confidence Intervals for AUC were also calculated, using the approach described previously by Hanley and McNeil.The validation dataset included 230 034 infants born to non-immigrant mothers and 70 098 born to immigrant mothers. The five most common countries of maternal birth were India (n=10 038), China (n=7468), Pakistan (n=5824), The Philippines (n=5441) and Vietnam (n=1408). The most common countries from the African continent were Somalia (n=833) and Nigeria (n=800), respectively. Descriptive characteristics of the validation cohort are provided in Overall model performance characteristics for non-immigrant populations and the top eight landed immigrant subgroups are presented in Model performance among preterm infants is provided in Our gestational age estimation model was able to discriminate between dichotomous preterm birth categories of immigrant infants with robust precision. For discrimination of ≥37 vs <37 weeks gestation, AUC ranged from a 0.927 among infants of maternal Somalian descent to 0.964 among infants of maternal Nigerian heritage. Similarly, the model was able to discriminate well between >34 weeks and ≤34 weeks gestational age, with AUC ranging from 0.966 for Nigerian and Bangladeshi infants to 0.994 among Filipino infants.This study validates a postnatal gestational age estimation model in subgroups of infants born to immigrant mothers living in Ontario, Canada. This work demonstrates reasonable performance of our previously published model to determine gestational age in infants born to immigrants of diverse countries of origin. Our findings provide proof-of-principle that metabolic modelling strategies couldbe robust in a variety of international infant cohorts.The strength of our approach lies in our ability to use population-based datasets to aggregate health and administrative data with ample sample size to evaluate model performance metrics across infant subgroups. In lieu of self-reported race or ethnicity, which were unavailable for our analyses, use of the Canadian Landed Immigrant Database provided data on infants born to immigrant mothers from a diverse range of countries. We acknowledge that naturalised immigrants living in North America may not be representative of individuals living in the country of origin, however, either due to admixture  or environmental factors . Indeed, whether subtle variation in model performance in specific immigrant subgroups, such as those from Bangladesh and Somalia, indicate a degree of biological or environmental influence warrants further consideration.As we work towards evaluating our model in other infant populations, we must consider existing limitations. First, our original approach included a relatively small sample size of preterm infants for model development, which is reflected in decreased accuracy of the model among the most severely preterm infants.In summary, our results provide reassurance that an algorithm developed from an Ontario-based population performs consistently well across a range of ethnicities. Ultimately, further validation studies will be required to evaluate the performance of this and other postnatal dating models in infants born and living across a range of international settings to determine if a single global algorithm or multiple regional algorithms should be derived.


IL7Rα and DCLRE1C deficiency SCID. Lymphocyte subset analysis and TREC quantification in the peripheral blood appear to be sufficient for confirmation of typical and leaky SCID and ruling out false positive (FP) results. Detection of secondary targets (infants with non-SCID lymphopenia) did not significantly affect the management or outcomes of these infants in our cohort. In the general, non-immunodeficient population, TREC rises along with gestational age and birth weight, and is significantly higher in females and the firstborn of twin pairs. Low TREC correlates with both gestational age and birth weight in extremely premature newborns. Additionally, the rate of TREC increase per week consistently accelerates with gestational age. Together, these findings mandate a lower cutoff or a more lenient screening algorithm for extremely premature infants, in order to reduce the high rate of FPs within this group. A significant surge in TREC values was observed between 28 and 30 weeks of gestation, where median TREC copy numbers rise by 50% over 2 weeks. These findings suggest a maturational step in T cell development around week 29 gestation, and imply moderate to late preterms should be screened with the same cutoff as term infants. The SCID NBS program is still in its infancy, but is already bearing fruit in the early detection and improved outcomes of children with SCID in Israel and other countries.Severe combined immunodeficiency (SCID), the most severe form of T cell immunodeficiency, is detectable through quantification of T cell receptor excision circles (TRECs) in dried blood spots obtained at birth. Herein, we describe the results of the first year of the Israeli SCID newborn screening (NBS) program. This important, life-saving screening test is available at no cost for every newborn in Israel. Eight SCID patients were diagnosed through the NBS program in its first year, revealing an incidence of 1:22,500 births in the Israeli population. Consanguine marriages and Muslim ethnic origin were found to be a risk factor in affected newborns, and a founder effect was detected for both  The purpose of newborn screening (NBS) is to identify newborns that are at risk for a panel of metabolic, endocrine, or hematologic diseases in which early diagnosis and prompt treatment will dramatically change disease outcome of affected patients . One sucn = 93). Finally, analysis was performed for 177,277 newborns.Data comprised of TREC results and accompanying epidemiologic parameters  for newborns born in Israel between October 1, 2015 and September 30, 2016, drawn from the computer archives of the Israeli National Center for NBS. Data entries with missing information or containing apparent typing errors were removed from analysis, as were entries for samples with poor DNA amplification (beta-actin <16 copies). Additionally, gestational age/birth weight norms were calculated and newborns for which birth weight was above 5 SD or below −5 SD for their respective gestational age were removed from analysis  or leaky SCID (T lymphopenia but >300/μl CD3+ T cells). The secondary target was to identify newborns with non-SCID lymphopenia . False positives (FPs) were defined as newborns with consecutive positive screening results, whose clinical presentation was unremarkable and immunological workup was negative for lymphopenia of any etiology.The primary target of the screening program for newborns was to identify those with SCID . Briefly, Dried Blood Spot (DBS) punches of 1.5 mm diameter are inserted into a black, 96-well PCR plate. DNA is eluted without extraction. Next, PCR amplification of TREC and beta-actin, an internal control for each specimen, is performed. Four PCR plates are processed in parallel, each plate containing a standard TREC curve in triplicates, as well as positive and negative controls for both targets.Severe combined immunodeficiency mutations were identified using either Whole Exome Sequencing (WES) or direct Sanger sequencing . Results were than validated through Sanger sequencing of patients’ parents, healthy siblings, or both.Cell surface markers of peripheral blood mononuclear cells (PBMCs) were determined by immunofluorescent staining and flow cytometry  using anti-CD3, anti-CD19, anti-CD16, and anti-CD56 from BD Biosciences; and anti-CD4 and anti-CD8 from Beckman Coulter.3-thymidine uptake assays (1 μCi/well) by culturing 105 PBMCs with phytohemagglutinin  or plastic bound anti-CD3 (10 µg/ml UCHT-1 from BD) for 72 h.T cell proliferation was tested by standard HSurface expression of individual TCRVβ families was analyzed using flow cytometry and a set of Vβ-specific fluorochrome-labeled monoclonal antibodies as previously described . Normal 3–106 copies of a plasmid with known TREC copy numbers. Patient and control samples were tested in triplicate, and the number of TRECs in a given sample was calculated by comparing the obtained cycle threshold value of the sample to the standard curve using an absolute quantification algorithm.T cell receptor excision circle copy numbers were determined using quantitative real-time PCR (qRQ-PCR). PCRs were performed as previously described  using asU test and the Kruskal–Wallis test were used to compare continuous variables between groups. Correlation between continuous variables was assessed using the Spearman’s rank correlation coefficient. All statistical tests were two-tailed. Differences were considered statistically significant when the p value was less than 0.05. Whenever applicable, results were presented as median, with 25th to 75th percentiles in brackets.Statistical analyses were performed using SPSS software . Given the skewed distribution of the data, the Mann–Whitney A total of 188,162 newborns were screened between October 1, 2015 and September 30, 2016. After subtracting entries with missing data, errors, and poor DNA amplification, data of a total of 177,277 newborns were analyzed. Of these, 51.5% were male. Median TREC copy number/blood spot for all DBS samples was 107 , median gestational age 39 weeks, median birth weight 3,240 g . 12,880 (7.26%) infants were born prematurely (<37 weeks gestation). 11,316  were moderate to late preterm , 1,126 (8.7%) very preterm (28 to <32 weeks), and 438 (3.4%) extremely preterm (<28 weeks). 1,614 (0.9%) were born Very Low Birth Weight , 12,293 (6.9%) were born Low Birth Weight .T cell receptor excision circle cutoff for retesting was initially set at 36 copies per blood spot and was gradually lowered to 23 copies per blood spot by year’s end. The Israeli SCID screening algorithm and the rate of positive results with different cutoffs are reviewed elsewhere (+ T cells (more or less than 300 cells/μl). Four patients had SCID variants with normal B cell counts (B+ SCID) and four had SCID variants with decreased B cell counts (B− SCID). All had normal numbers of NK cells. T cell proliferation was relatively normal in three patients and reduced or absent in the rest . Similarly, assaying TCRVβ repertoire did not prove sufficient for diagnosis as three of the SCID patients had only mildly abnormal results. The remaining patients had either a skewed repertoire (two patients) or the test could not be performed due to absence of T cells (three patients). TREC in DBS, obtained during the NBS, was undetectable in all SCID patients and in 7/11 infants in the FP group. The remaining four FP had an average of 10.4 ± 8.6 copies per blood spot. TREC quantification in peripheral blood was significantly reduced in all SCID patients compared to the FP group . Test sensitivity and specificity were calculated for each confirmation test with regards to its ability to distinguish between SCID patients and all other newborns referred to confirmation testing (FP and secondary targets). TREC in peripheral blood, total lymphocytes, and CD3+ are more sensitive and less specific than CD4+, TCRVβ, or proliferation assay  SCID. Six of the seven identified mutations were homozygous, whereas P8 was found to harbor compound heterozygote mutations . Complete confirmation tests were performed in only five of these syndromic patients. In nine newborns, the abnormal results were attributed to extreme prematurity with slow recovery of the immune system. Complete confirmation tests were available in only five of them. Four patients were diagnosed with secondary T cell immunodeficiency (three cases of chylothorax and one case of an infant born to a mother who was treated with immunosuppressive agents during pregnancy). Finally, five newborns, for which some of confirmation tests were abnormal (thus excluding them as FP) could not be classified. By 1 year, all of these children had normal repeat workup. They required no medical intervention.p < 0.001) between gestational age and TREC only in extreme preterms, and no correlation between age and TREC for all other age groups. When looking at median TREC for age groups, due to the large sample size, there was a significant difference in median TREC values between each preterm group and term infants , the differences between term and extremely preterm, 49 (29–92), or very preterm, 88 (52–149), were both highly significant  and meaningful. When looking at each birth week separately, one can observe a steady increase in median TREC from week to week between 23 and 28 weeks gestation, followed by a surge in TREC values between 28 and 30 weeks gestation, when median TREC rises by 32 (a 150% increase), another period of incremental increases from week 30 on until reaching a plateau around gestational week 38 . There was great variability in TREC results within each birth week, resulting in a mild positive correlation . The correlation between birth weight and TREC was more pronounced for extremely premature newborns (0.309) than for newborns of other age groups (0.147 for very premature newborns and no correlation for late preterm and term infants). All correlations were statistically significant, p < 0.001.T cell receptor excision circle copy numbers rose consistently and significantly along with birth weight Figure . Newbornp = 7E−202). The groups were similar in terms of gestational age  and birth weight . Correlations between gestational age and TREC, as reported above, were maintained when stratified into female  and male  infants. Correlations between birth weight and TREC, similarly, were maintained when sectioning the data into female  and male  infants.Median TREC for female neonates was 115.2 (74.6–179.7), compared to 100.8 (65.4–158.6) for male neonates. The difference was extremely significant (R2 = 0.9229). These data are limited by the fact that it pertains to a hospitalized population and may not represent the general population.Per the Israeli SCID screening algorithm, repeat DBS were collected and tested for TREC for preterm infants who remain hospitalized for an extended time period, regardless of their initial TREC result. We examined the rate of TREC increase per week for 4,212 preterm infants, for whom TREC results from multiple time points were available. Individual TREC dynamics were extremely variable. For all preterm infants, the median rate of TREC increase per week was 24 (−7.6 to +79). However, TREC increase rate changes dramatically with birth week Figure . For infp = 0.003). The correlation between TREC values of twin sets was low , significantly lower than the median TREC for singletons, as expected due to lower gestational ages and birth weight. Despite similar average birth weights between twins , the first twin in each set had a median TREC of 107 (67–177), compared to only 100 (62–166) for the second twin  were of satisfactory specificity to be used as standalone confirmation tests. In our cohort, lack of CD20+cells was also very informative to identify an underlying genetic defect that is associated with such immune phenotype (such as in DCLRE1C). It is important to remember, also, that TCR repertoire analysis could be of great value when lymphocyte profiling is inconclusive and clinical entities such as Omenn’s syndrome or SCID with maternal engraftment are suspected . Since the TCR repertoire test is not routinely available in many labs and the TREC assay by itself is known to be very sensitive also in diagnosing patients with these clinical phenotypes . This is compounded further by a slower recovery rate in preterms, with infants born 26 weeks gestation or younger in our cohort exhibiting a recovery rate of below four TREC copies/blood spot per week. Thus, for extreme preterms, a lower cutoff or a lenient confirmation approach (whereby extreme preterms are referred to confirmation only if still positive at 37 weeks corrected age), is acceptable (In both our cohort and previously published results by Barbaro et al., TREC copy numbers/blood spot rise significantly over a 2-week period between 28 and 30 weeks gestation . Though Further studies and analysis of relation between gestational age, birth weight, and TREC are required. As data from similar studies are aggregated, it may be possible to comprise a normal distribution chart for TREC based on gestational age and birth weight, in order to minimize FP results while allowing for prompt assessment of true positives.The purpose of NBS is to enable early diagnosis and treatment of life-threatening conditions. As reported above, seven of the eight SCID patients diagnosed through the Israeli screening program in its first year are currently alive and well, and are either post- or awaiting BMT. Before the initiation of the NBS program, such positive clinical outcomes were impossible for children with SCID.ER analyzed the data and wrote the paper; AL and AS performed all confirmatory and genetic tests and analyzed the results; SD, TS-L, and SA performed the screening for TREC and beta-actin and collected the results; TS, AB, AN, NM, OT, SH, ID, PS, and RS followed the patients; AE and RS analyzed the immunological results and supervised the writing of the paper. All authors approved the final version for submission and are accountable for all aspects of the work.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.