Objective: During bacterial vaginosis, an unexplained decrease of vaginal lactobacilli occurs. To identify whether these lactobacilli could be infected by phages, we isolated phages from vaginal lactobacilli and analyzed their potential virulence in attacking vaginal lactobacilli in vitro.					Methods: Vaginal samples were obtained from 39 reproductive-aged women. The selective Rogosa SL agar was used to isolate lactobacilli, from which phages induced by mitomycin C or released spontaneoulsy were analyzed by the agar spot method.					Results: Of 20 samples from women with vaginal infections, 12 did not have lactobacilli. From the remaining 8 infection samples and the 19 samples from healthy women, 37 Lactobacillus strains were isolated, from which 7 temperate phages were identified. Upon analysis, all 7 phages infected vaginal lactobacilli from the same and/or different women in vitro. Two phages, Φkc005 and Φkc007, had a broad host range, infecting 7 of 8 species tested. A control intestinal Lactobacillus phage also lysed several vaginal strains. One vaginal phage, Φkc039, was apparently lytic against vaginal lactobacilli from 7 other women. This phage was characterized as follows: plaque morphology, small and clear; burst size, 300 phages per cell; spontaneous induction rate, 1 per 106 cells; DNA, double-stranded and linear, 41 kb; and shape, a hexogonal head and a non-contractile tail.					Conclusions: Bacteriophages were isolated from vaginal lactobacilli of some women and were shown in vitro to lyse vaginal Lactobacillus strains from the same and/or different women. It was suggested that vaginal lactobacilli might be suppressed by phages.


A-R) has been reported to be important for palate formation, anesthetic action, and normal nervous system function. This subunit has also been implicated in the pathogenesis of Angelman syndrome and autism spectrum disorder. To further investigate involvement of this subunit, we previously produced mice with a global knockout of β3. However, developmental abnormalities, compensation, reduced viability, and numerous behavioral abnormalities limited the usefulness of that murine model. To overcome many of these limitations, a mouse line with a conditionally inactivated β3 gene was engineered.The β3 subunit of the γ-aminobutyric acid type A receptor . Crossing the floxed β3 mice to a cre general deleter mouse line reproduced the phenotype of the previously described global knockout. Pan-neuronal knockout of β3 was achieved by crossing floxed β3 mice to Synapsin I-cre transgenic mice. Palate development was normal in pan-neuronal β3 knockouts but ~61% died as neonates. Survivors were overtly normal, fertile, and were less sensitive to etomidate. Forebrain selective knockout of β3 was achieved using α CamKII-cre transgenic mice. Palate development was normal in forebrain selective β3 knockout mice. These knockouts survived the neonatal period, but ~30% died between 15–25 days of age. Survivors had reduced reproductive fitness, reduced sensitivity to etomidate, were hyperactive, and some became obese.A-R subunit. The floxed β3 knockout mice described here will be very useful for conditional knockout studies to further investigate the role of the β3 subunit in development, ethanol and anesthetic action, normal physiology, and pathophysiologic processes.Conditional inactivation of the β3 gene revealed novel insight into the function of this GABA A-Rs), mediates the bulk of rapid inhibitory neurotransmission in the adult mammalian central nervous system. GABAA-Rs are pentameric chloride channels assembled from 19 different subunits, α 1–6, β1–3, γ1–3, δ, ε, π, ρ1–3, and θ, 10 of the 11 offspring died 2 days postnatally and the survivor at day 21. Fifty-four percent (6 of 11) of these mice had cleft palate. These results were consistent with the phenotype of global β3 knockouts that were produced by traditional gene targeting [β3use line  to recom11 of 61 % vs expeFloxed β3 mice were also crossed to a synapsin I-cre (Syn-cre) transgenic mouse line ,24 to prF/F, cre- by β3F/F, cre+ mating pairs for genotype analysis. This study revealed that β3F/F, cre+ pups were present at a frequency that was similar to β3F/F, cre- pups . Visual analysis of the palates of a subset of these mice revealed that 0 of 9 β3F/F, cre+ pups examined in detail had cleft palate. Thus, we conclude that pan-neuronal knockout of β3 is compatible with in utero survival and normal palate development.To investigate the effects of embryonic, pan-neuronal knockout of β3, pups were collected before birth (embryonic day 17.5–20.5) from β3F/F, cre- by β3F/F, cre+ mating pairs was performed. This study revealed that only ~28% (108/384) of the mice genotyped were cre+, in contrast to the ~72% (276/384) that were cre-. This differs from the expected 1:1 ratio. These results suggest that ~61% of the β3F/F, cre+ mice died before weaning. Many of the cre+ pups died within a few days of birth. Thus, we conclude that pan-neuronal inactivation of β3 by Syn-cre appears to result in a variably penetrant, non-cleft palate, neonatal lethality.To investigate the effects of pan-neuronal knockout of β3 on postnatal survival and development, a genotypic analysis of weanling mice derived from β3F/F, cre+ mice that survived beyond weaning were observed while in their home cage and during handling. The cre+ mice were overtly indistinguishable from control, cre- littermates. Unlike the global β3 knockouts [F/F, cre+ mice were found to be fertile and females displayed normal maternal care.Gross behavior of the β3nockouts , these cTo examine the degree to which the β3 gene product was reduced in various brain regions of the knockouts, western blot analysis of hippocampus, cortex, and cerebellum was performed using a β3 specific antibody. As shown in Figure Pan-neuronal knockout mice were tested for sensitivity to the intravenous anesthetic etomidate and ethanol using the standard loss of the righting reflex (LORR) assay. For etomidate, the analysis revealed a significant effect of gender (p < 0.01). Therefore, males and females were analyzed separately. This gender difference was also seen when etomidate was administered intravenously (data not shown). As shown in Figure To produce a β3 knockout that is restricted only to postnatal forebrain neurons, the floxed β3 mice were crossed with an α CamKII-cre mouse line . This moF/F, cre- by β3F/F, cre+ produced normal sized litters with no apparent neonatal mortality. However, some cre+ mice died between 15 and 25 days after birth, a time frame that coincides with expression of cre recombinase from the α CamKII-cre mouse line [Mating of β3To examine the impact of α CamKII-cre mediated recombination on the β3 gene product, western blot analysis was conducted on hippocampus, cortex, cerebellum, and hypothalamus using a β3 specific antibody. As shown in Figure F/F, cre+ mice was tested by crossing to wild type mates. Both males and females were reproductively impaired. Two of six females that were test-mated consistently produced and cared for normal sized litters. The four other females either did not produce any offspring or produced litters infrequently and usually failed to care for pups. Male β3F/F, cre+ mice were test mated to superovulated females. None of the four cre+ mice produced copulation plugs, in contrast to 12 of 12 cre- mice that were tested for comparison. In addition, these four cre+ mice were housed with β3F/+, cre- females for 2–4.5 months. One male sired a seemingly normal number of litters during that time (n = 4), whereas the other three males each only sired a single litter. Thus, male reproductive fitness also seemed to be compromised.Reproductive fitness of β3F/F, cre+ became obese in adulthood. To quantify this, the body weights of mice from 4–16 weeks of age were measured. As shown in Figure During the course of these studies, we also noticed that some, but not all, β3To find out if the obese mice also had a greater food intake, food consumption between knockout males and controls was compared. By 20 weeks of age, all surviving cre+ males were obese . Average food consumption assessed over the course of 8 days was greater in knockouts compared to controls  of cleft palate. This result agrees with studies that have reported a high frequency of cleft palate in mice that lack β3 due to traditional knockout of β3  or radiaA-Rs in nonneuronal tissues is critical for normal development of the palate. An alternative explanation is that the time course and/or the extent of the genetic modification in the present study or in the study of Hagiwara et. al [While these global knockout/transgenic studies clearly define a role for β3 in normal palate formation, they do not distinguish between a neuronal and nonneuronal site of action. To dissect the contribution of neuronal vs nonneuronal β3 in palate formation, a conditional knockout mouse line which lacked β3 in most neurons was developed. In these pan-neuronal knockout mice, cre-mediated recombination was under the control of the rat synapsin I promoter. Although the time course of cre-mediated inactivation of β3 in these mice was not examined in detail, this same Syn-cre transgenic mouse line has been demonstrated to induce recombination primarily in neurons of the central nervous system starting at approximately embryonic day 12.5 ,24. Clefa et. al  was not a et. al .in utero but is required for the transition to life immediately after birth. A high level (~61%) of perinatal mortality in pan-neuronal β3 knockouts was observed. The fact that pan-neuronal knockout did not result in cleft palate indicates that the perinatal mortality is not due to a defect in the palate that prevents feeding. The cause of death in these animals has not been determined. Furthermore, forebrain selective deletion of β3 (initiated ~2 weeks after birth) rescues the mice from perinatal mortality, but it fails to ensure survival to adulthood. Approximately 30% of forebrain selective β3 knockout mice die between 2 and 4 weeks of age. These results reveal that neuronal β3 is required beyond the perinatal period for normal viability. Furthermore, these results confirm that the mortality is not due to aberrant embryonic development. Instead, we conclude that there is a role for β3 in neuronal function that is required for survival.The results presented here also shed new light on the requirement of β3 for normal viability. Global deletion of β3 resulted in a very high frequency of neonatal mortality, a result that is consistent with earlier studies of global β3 knockouts ,27. ThesThose conditional β3 knockouts that survived to adulthood displayed a variety of abnormalities that varied considerably between animals. Forebrain selective deletion of β3 also resulted in deficits in reproductive fitness, maternal care, and overt behavior. These phenotypes were similar to that observed in global β3 knockouts ,27.Hyperactivity is a common observation in global β3 knockout mice . AlthougA-R agonists to be stimulators of food intake [A-R response that increased the functional activity of GABAA-Rs in brain areas involved in regulating food intake, e.g., the hypothalmus. However, the role of GABA in the mechanisms governing food intake and body weight is unclear. GABA has been shown to be colocalized with neuropeptide-Y neurons in the arcuate nucleus of the hypothalamus [A-R-containing neurons in the lateral parabrachial nucleus of the brainstem have also been postulated to be involved in GABA-mediated increases in food intake [A-R subunit in an anatomically-specific manner is an important first step in unraveling the role of GABA in regulating body weight.Body weights were increased by forebrain selective knockout of the β3 subunit in both male and female mice Figure . This mad intake , the incthalamus  which cothalamus  that supd intake . While GA previous study revealed that global β3 knockout mice were less sensitive to the sedative/hypnotic effects of etomidate . HoweverA-Rs are also important for mediating the effects of ethanol [Reportedly, β3 containing GABA ethanol . However ethanol  of the βIt is unknown why the cleft palate, mortality, body weight, and behavioral phenotypes are variably penetrant within the various β3 mutant mouse lines. It is possible that animal-to-animal variability exists because of genetic heterogeneity in the background of the animals. None of the mutant mouse lines used here were maintained on a uniform, inbred genetic background. Genetic background has been demonstrated to substantially alter the incidence of cleft palate in global β3 knockouts . DetaileA-R was also reduced   to produMice harboring the neo-containing floxed β3 locus were crossed to an actin-FLPe deleter mouse line  to permaA-R α1 using an α1 specific antibody . Primary antibodies were detected with HRP-conjugated goat anti-rabbit antibody  and visualized by chemiluminesence . Immunoreactive protein levels were compared by densitometric measurement of band intensities and analyzed using Student's t test.Protein levels in various brain regions of pan-neuronal and forebrain selective knockouts were analyzed using semi-quantitative Western blotting as described . BrieflyMice were individually weighed at weekly intervals between 4 and 16 weeks of age. Data were analyzed by repeated measures ANOVA.Food consumption was determined on singly-housed 20 week-old male mice by weighing the amount of food consumed during an eight day period. The average daily food consumption (in grams) was calculated and data were analyzed using a Student's t test.Adult forebrain selective and pan-neuronal knockout and control mice were tested for sensitivity to ethanol or the anesthetic etomidate using a standard sleep time assay as described . BrieflyOperant learning was assessed using a step-through passive avoidance task as previously described in detail . Brieflyt-test using the software program PRISM 4 .Spontaneous locomotor activity was assessed using an automated open field test apparatus. Mice were allowed to acclimate to the test room for 30 min prior to testing. Mice were placed individually into clear plastic monitoring chambers measuring 72 × 32 × 32 cm. Locomotor activity was measured via seven sets of photoelectric sensors evenly spaced along the length of the monitoring chamber, 4 cm above the floor of the chamber . Total activity was recorded in arbitrary units reflective of the number of times a mouse interrupted the photoelectric sensors during a 10-min monitoring session. Data was automatically recorded and stored by computer. Data was collected in two bins, one representing the total activity and the other representing the ambulatory component of total activity. Total activity includes ambulatory activity as well as locomotor activities directed towards rearing, grooming and exploring small regions of the testing chamber. Data were analyzed by an unpaired Student's The author(s) declares that there are no competing interests.CF and SLH performed experiments and analyzed data; DFW performed some western blot experiments and analyzed data; SMH directed body weight and food consumption studies; TMD performed and analyzed locomotor behavioral experiments; GEH conceived of the study, performed and/or directed experiments, analyzed data, and wrote the manuscript. All authors helped draft the manuscript and have read and approved the final version.


Am J Trop Med Hyg 83:56–60, one of the author's names appears incorrectly. The correct name is Panagiotis Karanis, not Karanis Panagiotis. The journal regrets the error.In Am J Trop Med Hyg 83:174–182, there were errors in In Correction: Results, “Overall Incidence”, 3rd paragraph, last sentence: “With the exception of the Pacific and Middle Atlantic regions, incidence in all regions increased more than 100% from the first half (2000–2003) to the second half (2004–2007) of the reporting period (g period .”


Am J Trop Med Hyg 81(6): 927–933 by Yadon and Schmunis, there is an error in reference 3. The correct reference is “Carlier Y, Torrico F, 2003. Congenital infection with Trypanosoma cruzi: from mechanisms of transmission to strategies for diagnosis and control. Revista da Sociedade Brasileira de Medicina Tropical 36: 767–771.” The journal regrets the error.In


Structural variations in the form of DNA insertions and deletions are an important aspect of human genetics and especially relevant to medical disorders. Investigations have shown that such events can be detected via tell-tale discrepancies in the aligned lengths of paired-end DNA sequencing reads. Quantitative aspects underlying this method remain poorly understood, despite its importance and conceptual simplicity. We report the statistical theory characterizing the length-discrepancy scheme for Gaussian libraries, including coverage-related effects that preceding models are unable to account for.Deletion and insertion statistics both depend heavily on physical coverage, but otherwise differ dramatically, refuting a commonly held doctrine of symmetry. Specifically, coverage restrictions render insertions much more difficult to capture. Increased read length has the counterintuitive effect of worsening insertion detection characteristics of short inserts. Variance in library insert length is also a critical factor here and should be minimized to the greatest degree possible. Conversely, no significant improvement would be realized in lowering fosmid variances beyond current levels. Detection power is examined under a straightforward alternative hypothesis and found to be generally acceptable. We also consider the proposition of characterizing variation over the entire spectrum of variant sizes under constant risk of false-positive errors. At 1% risk, many designs will leave a significant gap in the 100 to 200 bp neighborhood, requiring unacceptably high redundancies to compensate. We show that a few modifications largely close this gap and we give a few examples of feasible spectrum-covering designs.The theory resolves several outstanding issues and furnishes a general methodology for designing future projects from the standpoint of a spectrum-wide constant risk. The relevance of genomic structural variation (SV) to human medical disorders is well-known ,2 and efet al. [Broadly speaking, SV encompasses translocations, inversions, and copy number variations and other types of inserted and deleted sequences (indels). Here, we focus on the last category, which is believed to occur the most frequently . Historiet al.  proposedα < 1%. Other procedures call for considering inserts more than 2 deviations from the average [Although investigators are actively pursuing this technique -12, it i average .α and β  distributed ,27 as(1)λ and σ are the average and standard deviation of length, respectively, and exp is the exponential function. End-pair alignment of randomly selected inserts yields a sample whose members span a candidate SV region having a magnitude at least δ = |x2 - x1|  spans a site of SV. Let Hm and Ht represent the status of the SV as homozygous and heterozygous, respectively. If the genomic source has a haploid length G, the Bernoulli probabilities of S areConsider the event Wherel ≥ m (Eq. 11).is the minimum admissible insert size and m would actually be smaller than the expression given in Eq. 5, however, there is no reliable basis for measuring insert length in such cases, even for perfect, un-ambiguous alignments [m = 2r for these cases.If singleton reads are allowed for ISV, then ignments . Consequignments ,10,11,15tributed ,27 with whereN is the number of inserts that have been processed.is the general rate expression and α and β, respectively [h0, as the absence of SV of size at least δ.The classification problem for SV is characterized by the probabilities of incorrectly calling out an instance of SV  or overlooking a true instance of SV . Such scenarios are described by the traditional inference probabilities ectively , which cC be the event that the sample average length falls within some specified confidence interval, thenLet h0 and these are significant at a level of α = 1 - P(C). This statement immediately implies the following corollaries.is the probability that differences between the sample and library average lengths are attributable to the sampling effect . InstancC represents intervals λ - δ ≤ ℳ and ℳ ≤ λ + δ for ISV and DSV, respectively, where the variable m embedded in these equations assumes appropriate values, as given by Eq. 5. Homozygous and heterozygous configurations are differentiated according to Eq. 6.The standard model recognizes the inherent asymmetry between ISV and DSV, employing separate one-tailed tests for each. Specifically C is the interval λ - δ ≤ ℳ ≤ λ + δ and where m = 2r for both ISV and DSV. Homozygous and heterozygous configurations are again differentiated according to Eq. 6. This model can be regarded as a direct extension of previous work [The symmetric model disregards the asymmetry between ISV and DSV and implements classification according to the two-tailed test, where ous work ,11,15, tD be the event that an instance of SV is provisionally detected (i.e. not overlooked), as defined under the following conditions: It is spanned by at least one insert and the length of at least one of those spanners is different from the library mean by a specified threshold amount τ. We haveLet t = τ/(σ). Consequently, the detection power P(D), sometimes written more traditionally as 1 - β, is given by Eq. 8, where C is replaced by D.where Here, we examine some of the consequent properties of the so-called length-discrepancy method that will have implications for future projects. This discussion is framed in terms of several insert types that are in either experimental or common use Table . We concσ, to control false-positive errors [et al. [δ ≥ 8 kb using fosmids. Fig. Thus far, concerns have predominantly focused on the rate of false-positive SV declarations. The general methodology has been one of assuming symmetric behavior of ISV and DSV and subsequently employing elementary Gaussian thresholds, usually ± 3 e errors ,9-11,15. [et al.  aimed toδmin, for each specific case. For example, the significance level α = 1% for fosmids is realized for roughly all δ ≥ 8 kb at ρ = 8× physical coverage. While matching the δmin = 8 kb calculated by Tuzun et al. [α is actually highly dependent upon the physical coverage, as shown in the figure, while δmin is only weakly dependent upon ρ.The asymmetry between ISV and DSV is quite clear. Statistical significance of DSV detection is basically governed by a single minimum size, n et al. , this daδmin, but also a maximum size, δmax. The latter arises because placement constraints for fixed λ increase with ISV size. Physical coverage is again tightly coupled with these limits. Increased ρ not only renders the better (lower) values of α more accessible, but also widens the range of acceptable δ. For example, heterozygous ISV using fosmids does not register at the 1% significance level even at 8× physical coverage. However, at 12×, events are significant roughly in the range 3,700 ≤ δ ≤ 12,700 and at 16× in 2,800 ≤ δ ≤ 19,200. The upper limit on δ increases somewhat faster with redundancy than does the lower limit for ISV.The situation for ISV is quite different in that there is not only a r/λ = 20% here vs. about 1.5% for fosmids. Increased variance obviously degrades the statistics, but ironically, so do "better" read lengths. The latter phenomenon arises because of the requirement that reads lie outside the insertion, which implies fewer placement possibilities for covering (Eqs. 4 and 5). These observations indicate that short-read data will generally have to be generated at much deeper redundancies than large-insert clones (discussed in more detail below) and also raise the issue of optimal read lengths. However, the latter depend on all the methods one might use to find variation, so it cannot be settled on the basis of the length-discrepancy approach alone.While trends are similar for Illumina GA short inserts, there are notable differences with respect to physical coverage. In particular, ISV at 1% significance does not become feasible until roughly 24×, and even then its range is quite small. The main parametric differences from fosmids are that the variance is higher here, 12% coefficient of variation (COV) vs. 7% reported in ref. , and thaα = 50% as δ increases as a consequence of vanishing covering probabilities. Inferences in these regions are no better than a coin flip. Notice that the fosmid curves also approach the 50% asymptote as δ → 0. The underlying factor here is a vanishing precision, not unlike that experienced when evaluating the small difference between two increasingly large numbers.All ISV curves approach the asymptote ad hoc alternatives to the null hypothesis [τ. For instance, for ISV, at least one covering insert must be shorter than λ - τ, meaning its aligned length will be less than λ - τ - δ.The concept of detection power is never entirely precise because it requires adoption of pothesis . Ours reτ = 0 for heterozygous ISV. Each curve is asymptotic for δ ≪ λ, but rapidly decays as SV size approaches the insert size.Thresholds can be specified in numerous functional ways, each of which has certain implications for finding SV of different sizes. Fig. Asymptotics are readily shown as a special case of the model to beλ, are penalized, but increased redundancy can compensate for the shortfall. For instance, asymptotic power for Illumina GA data could be made equal to that for fosmids if its redundancy were roughly 60% higher than fosmid redundancy. Note the trend for each insert's power curve to more visibly resemble a unit step function as redundancy is increased. In comparing α to power, it appears the latter is quite acceptable. For example, 16× fosmids are upper-bounded at roughly 20 kb for α = 1% significance  power for DSV. Here again, we see that inserts having relatively long reads, i.e. larger r/ls - δ, will be known, although ls and δ themselves generally are not. However, one can readily quantify power versus the aligned length of the average insert, for example using some τ = δ/C. This information might serve as a basis for constructing detection rules by correlating a priori known theoretical entities with other heuristic information.Defining rigorous detection rules is a challenging task for algorithm developers. Again using ISV as an example, the aligned length of the shortest covering insert, α is largely constant for COV ≤ 7% over a wide range of redundancies, implying that special efforts aimed at further reducing fosmid library variance would be unwarranted. While some sensitivity is actually realized for very small ISV, i.e. less than 10% of insert size, the limit on precision mentioned above renders these instances irrelevant.Library variance is conventionally thought of as something that should be minimized to the greatest extent possible in order to improve SV detection . This viσ and λ - m, respectively. The second term represents the familiar "Lander-Waterman" type of covering mechanism which should ideally provide the bulk contribution, but its potency drops considerably for larger δ via m (Eq. 5). This reflects the simple fact that the average insert will not cover very well under these circumstances. Performance can be recovered in a seemingly counter-intuitive way by increasing the library variance, which implies there are more longer-than-average inserts in the library. Fig. Another curious phenomenon associated with ISV lurks in Eq. 7. Its exponential and error function terms have leading coefficients α rapidly becomes responsive to library variance over slight changes in m/λ , meaning that there would be some level of improvement if library standard deviation could be reduced. However, for a small decrease to m/λ = 0.5 (25 bp ISV), the situation worsens in two ways. Not only does α become appreciably more sensitive, but attempts to compensate with increased redundancy are less effective. For example, at 12% COV we could reduce α by 68% (from 11% to 3.5%) by doubling ρ from 12× to 24×. If Illumina GA libraries were improved to fosmid-level 7% COV, we would instead see α reduced by 90%, from 4.3% to 0.44%. Yet, the curves show that still more resolution could be wrung out, all the way down to about 4% COV (σ = 10 bp). This overall behavior is again a consequence of the relatively long read lengths, which drive down δ for a given m/λ ratio. Although these observations suggest investing more effort into reducing σ for Illumina GA libraries, the balance against economic considerations has not been conclusively established.The situation is appreciably different for Illumina GA inserts, where m/λ Fig. . It is mpriori specify the combination of insert types and corresponding redundancies that will best characterize the relevant fraction of the spectrum [δ ≤ 20,000 bp.  Investα, in this case 1% [Fig.  case 1% ,10,11,15 case 1% .) The daMedium and large Illumina GA inserts, combined with fosmids, readily handle SV above about 300 bp using assorted physical coverages in the 15× – 20× range. Although the small insert library does cover much of the neighborhood on the lower end, there is a conspicuous gap between 100 and 200 bp, precisely where many variants could be expected [Let us assume hypothetical Illumina GA libraries whose COVs are reduced by one-third of their conventional values. Also, replace the 3 kb Illumina GA library with a 1.5 kb GA library. These two modifications largely erase the gap, i.e. curves for adjacent libraries now intersect at roughly 30×. Switching to the 1.5 kb library is primarily responsible for this closure, although reduced COVs further improve spectral coverage on the lower ends of the respective libraries. This effect is especially relevant to extending detection range for the smallest SVs.α. Fig. ρ = 30× for 250 bp and 1.5 kb GA libraries and ρ = 12× for 10 kb GA and 40 kb fosmid libraries, netting SV within approximately 20 ≤ δ ≤ 13000. The second prescribes 50× for the 250 bp GA library and 18× for the remaining 3 libraries, widening the range to roughly 13 ≤ δ ≤ 21000. Although many other designs are clearly possible, these illustrate some of the interesting trade-offs faced by the investigator. For example, maximally extending the range raises the possibility of including additional insert types, e.g. a BAG library. Costs are also vastly different over the various insert types, with small GA being the cheapest, larger GA being more costly because of library inefficiencies, and fosmids being the most expensive according to direct library and sequencing costs. However, these issues are tempered by the fact that long inserts efficiently leverage sequence redundancy, 2 ρ r/λ. For example, 20× fosmid physical coverage translates to only about 0.5× sequence redundancy [Spectral charts are useful for designing projects according to the requirement that no gaps remain in the SV detection spectrum for a desired priori, which may not be the case for tumors. Finally, we do not account for mapping or sequencing errors, library complexity, the ability of algorithms to distinguish between the reads covering both alleles of a heterozygous SV site [Our theory describes SV statistics in more general terms than currently available, though it still depends upon a number of idealizations for the sake of tractability. Specifically, the covering process is taken to be independently and identically distributed. We also presume genome size is known a  SV site , instancα, etc. It would be quite useful to optimize on the basis of total project cost, but variation in library production and sequencing costs among different insert types, lab environments, production methods, bio-economical feasibility of minimizing COV for inserts, etc. places this goal beyond the present scope. Results shown here suggest an intermediate number of libraries, i.e. three to five, will typically give the best results. Too few will incur extremely large per-library redundancies because of spectral gaps and the asymptotic nature of spectral coverage , then the mean length ℳ of random samples of size k  will be Gaussian with standard deviation σS = σ/If insert lengths are Gaussian distributed with standard deviation n σS = σ/. The pron σS = σ/a', b') =  for Eq. 2, but change to a' → - ∞ or b' → ∞, as appropriate, in Eq. 3. By making a change of variables, x = (l' - λ)/(σS), the theorem of Integration by Substitution [where (titution  can be iλ - δ is changed to -δ/(σS) and λ + δ to δ/(σS). All intervals follow directly by substituting σS = σ/x) = -erf (x) and erf (∞) = 1 [where erf is the Gaussian error function  and wher (∞) = 1 , the latG possible placements of any insert, since the haploid genome and its constituent chromosomes are very large. For homozygous SV, the apparent number of placements is then only G. For DSV, we disqualify cases where either read of an insert intersects an SV site, whereby the spanning event is realized only if the interior region between the paired reads contains l - 2r placements satisfying this condition, implying spanning probabilities of (l - 2r)/G and (l -2r)/(2G) for Hm and Ht, respectively. For ISV, the breakpoints are separated, and the variant exists as segment l - 2r - δ + 1, which is well-approximated as l - 2r - δ, given that l ≫ 1. Spanning probabilities for Hm and Ht follow by appropriate division. The fact that 2r and 2r + δ are the minimum admissible insert lengths for DSV and ISV, respectively, is a direct consequence of both the mathematical fact that probability values cannot be negative and the physical observation that these are the minimum values at which an insert could span the respective types of SV.Lemma 2 is proved by straightforward enumeration. First, there are roughly 2nt  Fig. . The numP(S) for each insert (Lemma 2), the covering process for a site of possible SV is clearly binomial, i.e. an individual insert either covers, or does not cover the site. For SV projects, the number of inserts processed, N, and P(S) are necessarily large and small relative to unity, respectively. The Poisson distribution for P (k) then follows directly from the standard binomial limiting argument [μ = N·P(S) is the Poisson rate. We will demonstrate only the homozygous configuration, since the heterozygous case follows along exactly the same arguments.Based on a Bernoulli probability argument , where μS, whereby an insert covers a possible site of SV, is conditioned upon its length. Consequently, it can be expressed according to the Law of Total Probability [The prior Bernoulli probability of event bability  as(11)P (ℒ = l) and P(S|ℒ = l ∩ Hm) have already been established by Eq. 1 and Lemma 2, respectively. The latter also establishes m as the lower bound, while the upper bound is a consequence of the observation that inserts will not exceed the subject haploid DNA length. Biological constraints actually restrict inserts to much smaller sizes. However, it will be clear momentarily that the actual value of the upper bound is not particularly important because the associated functions rapidly converge to their respective limits.where the summation appears by virtue of the events being mutually exclusive. Note that P(S) is further developed by first transforming to an integral representation. Substituting P (ℒ = l) and P(S|ℒ = l ∩ Hm) and moving constants outside the integral, we findThe form of where the error of transformation can be shown by the Euler-MacLaurin Theorem  to beλ ≫ 1, even for inserts considered to be fairly short, and likewise σ ≫ 1, the magnitude of ϵ will be acceptably small for cases of practical interest. Making a change of variables similar to that shown in the proof of Theorem 1 the integral evaluates toBecause G is extremely large (order 109 for the human genome), the second exponential term vanishes, while the first error function term is asymptotically equivalent to unity.  Applying the identity erf (-x) = -erf (x) [P(S), from which μ = N·P(S) in Eqs. 6 and 7 immediately follow.Given that -erf (x)  to the sC whereby the average length of the sampled inserts falls within a certain confidence interval is conditioned upon how many inserts actually comprise the sample, whereby from the Law of Total Probability [The event bability  we findfrom which Eq. 8 follows directly.i whose length Li is within some range of τ isUsing the same integral methods shown in the proof for Thm. 1, the probability of selecting any single insert t = τ/(σ). Because inserts are presumed independent of one another, the probability that k inserts picked from the library are all within the specified range is P1 ∩ P2 ∩ ⋯ ∩ Pk = k inserts spanning the SV site, whereby application of the concept of Total Probability (similar to what is shown in proof of Thm. 4) yields P(D).where P (k), whose floating-point numerator and denominator will both tend to overflow as k grows large. In these cases, we evaluate P (k) in logarithmic form. Using Stirling's Series for the factorial term [The statistical properties of SV depend upon numerical evaluation of ial term we find that the expression can be writtenk. Given the definition of redundancy in Table N/G with ρ/λ in Eq. 7. Also, it will often be impractical to sum all N + 1 terms, e.g. in Eq. 8, so we use a simple convergence rule that halts the computation when the percentage change due to the current term falls below a small number, typically 10-8.for large Accurate numerical methods exist for evaluating the Gaussian error function , and sucSV: structural variation; COV: coefficient of variation.MCW conceived and constructed the mathematical theory and wrote the paper. Both authors approved the final manuscript.


Although the physiological effects of immunocontraceptive treatment with porcine zona pellucida (PZP) have been well studied, little is known about PZP's effects on the scheduling of reproductive cycling. Recent behavioral research has suggested that recipients of PZP extend the receptive breeding period into what is normally the non-breeding season.Equus caballus) living on Shackleford Banks, North Carolina for 4 years pre- and 8 years post-contraception management with PZP . Gestation lasts approximately 11–12 months in wild horses, placing conception at approximately 11.5 months prior to birth. Since the contraception program began in January 2000, foaling has occurred over a significantly broader range than it had before the contraception program. Foaling in PZP recipients (n = 45 births from 27 mares) has consistently occurred over a broader range than has foaling in non-recipients (n = 52 births from 19 mares). In addition, current recipients of PZP foaled later in the year than did prior recipient and non-recipient mares. Females receiving more consecutive PZP applications gave birth later in the season than did females receiving fewer applications. Finally, the efficacy of PZP declined with increasing consecutive applications before reaching 100% after five consecutive applications.To determine if this is the case, we compiled foaling data from wild horses (For a gregarious species such as the horse, the extension of reproductive cycling into the fall months has important social consequences, including decreased group stability and the extension of male reproductive behavior. In addition, reproductive cycling into the fall months could have long-term effects on foal survivorship. Managers should consider these factors before enacting immunocontraceptive programs in new populations. We suggest minor alterations to management strategies to help alleviate such unintended effects in new populations. The extirpation of predator species has resulted in the expansion of free-ranging ungulate populations in North America Equus caballus) has become a model for evaluating the effects of PZP, thanks chiefly to the seminal nine-year study of contracepted mares on Assateague Island National Seashore In recent decades, the wild horse , one to several female(s), and their offspring On Shackleford Banks, increased reproductive behavior in the post-breeding season by mares has resulted in increased male attentiveness Here we use the birth dates of foals to estimate dates of conception for PZP recipient and non-recipient mares on Shackleford Banks, North Carolina. Gestation length in wild horses is 11–12 months This study was conducted on Shackleford Banks, a barrier island located approximately 3 km off the coast of North Carolina, USA. The island was 15 km in length, and varied between 0.5 and 3 km in width. The horse population on Shackleford Banks has been co-managed by the National Park Service and the Foundation for Shackleford Horses since 1996.The reproductive units of Shackleford horses are typical of feral equids. They are coherent harem groups of one, or sometimes two or three stallion(s), one to several mare(s) and their offspring In January 2000, the National Park Service began the application of PZP for the purposes of immunocontraception. The National Park Service administers PZP in the spring (late February through April) each year. Mares receive their initial treatment at 1.5–2 years of age. Each injection includes 100 micrograms of PZP with an adjuvant (combined at the darting site). Initial doses include Freund's Complete Adjuvant, Modified, Mycobacterium butyricum . All succeeding doses include Freund's Incomplete Adjuvant (Sigma #F5506). In a given year, an average of 63% of all reproductive mares are inoculated with the vaccine (range = 37–88%). The authors of the present study are not and have never been in charge of making management decisions regarding this herd.The National Park Service began their management of the Shackleford Banks population prior to the use of PZP, performing five gathers between November 1996 and January 2000. During these gathers the majority of the population was rounded up and individuals testing positive for equine infectious anemia were removed and either euthanized or quarantined. As such forms of management could conceivably influence reproductive cycling, we analyzed foaling dates before and after gathers (but before contraception management) using a linear mixed effects model .From January 2000 – January 2008, 38 foals (conceived due to contraception failure or administration scheduling) were removed from the island for the purpose of population control. The majority of removals (92%) were conducted in the January following the foals' birth.We recorded foaling data before contraception management (1995–1997) during a study of mother-infant behavior We identified individual horses by color, sex, age, physical condition, and other distinguishing markings including freeze brands. We monitored pregnant mares 2–3 times per week to ensure accurate estimation of foaling date. Births are not commonly witnessed, as mares will generally give birth away from their harem group We designated births as coming from current recipient, prior recipient, or non-recipient mares. Current recipients received PZP treatment the year they conceived . Prior recipients received PZP treatment at some point earlier in their lifetime, but not the year of conception. Non-recipients had never received PZP at the time of conception.Physical condition is an important factor to consider when monitoring reproductive behaviors. Animals in better condition will have more resources to allocate to mating behavior and physiology than will animals in poorer condition. We assessed mare condition via rump scoring. We determined rump scores examining the curvature of the line between the tailbone and the point of the hip. Scores were based on a scale from 1 to 5; a score of 1 being the poorest We collected all climatology data from the Morehead City WNW Station at 34° 44′N; 76° 44′W, approximately 8 km from the study site. We obtained all data from the National Climate Data Center (2009) P-values.We analyzed data in R version 2.11.1  and JMP, version 7 . We used Fligner-Killeen tests to determine if variation in foaling dates differed before and after contraception management and between PZP recipients and non-recipients. This test is more conservative than Fisher's or Bartlett's tests and is less sensitive to outliers and non-normal data We used linear mixed effects models to analyze monthly weather patterns. These statistical models included year as a random effect and the following fixed effects: linear through fourth order polynomial functions of month , management regime (before or after contraception management), and two-way interactions with management regime. We selected best fit models using Akaike's Information Criterion adjusted for small sample size (AICc). Because temperature data were temporally autocorrelated, we included an autoregressive moving average correlation structure 2,84 = 2.03, P = 0.14), suggesting that age did not confound differences among groups. From the above mixed effects model, differences in foaling dates among current, prior, and non-recipients were calculated using an F-test for linear combinations To test for differences in foaling date with PZP treatment, while controlling for pseudo-replication and unequal variances, we used a linear mixed effects model. The model included mare ID as a random effect and a function allowing different variances across groups  P = 0.63). Additionally, the range of birth dates did not differ significantly between these groups . Finally, AICc from our analysis of foaling dates . After controlling for weather and mare age, mean foaling date did not differ before and after gathers  or the number of years between pregnancies  influenced foaling date, we ran separate linear mixed effects models using these as predictor variables. Each model used only data from PZP recipients and included mare ID as a random effect. To determine if results from this analysis could be influenced by year, we ran a linear mixed effects model on post-management, non-recipients predicting foaling date by year.P-values were >0.10.We analyzed PZP efficacy using a generalized linear mixed effects model with a binomial error distribution and mare ID as a random effect P<0.0001; pre-contraception: n = 65 births from 45 mares over 4 years; post-contraception: n = 97 births from 46 mares, over 8 years). In the eight years following contraception management, mares receiving PZP foaled over a broader range of months than did non-recipient mares .After contraception management, foaling occurred over a broader range of months than before contraception management : 40.79, generalized r2 = 0.48, P<0.001). Mares that had received PZP earlier in their lifetime, but not during the year of conception (prior recipients), gave birth 0.90 months later than pre-contraception mares on average . Non-recipient mares that never received PZP themselves, but gave birth after the general population was managed with PZP, gave birth 1.01 months later than pre-contraception mares on average . This translates into current PZP recipients giving birth 2.34 months later than non-recipient animals and 2.46 months later than prior recipients . Furthermore, birth dates were about 0.38 months later for each degree centigrade above average in the latter half of the breeding season and 0.05 months later for each centimeter of rain above average . In general, mares gave birth 0.05 months earlier for every year of age .On average, current PZP recipients gave birth 3.36 months later than did pre-contraception mares, according to our linear mixed effects model, which controlled for heterogeneity of variances, weather influences, and mare age . It is unlikely that these patterns resulted from a general trend across years, as birth dates among non-recipient animals did not change with calendar year .Among mares treated with PZP, those receiving a higher total number of applications foaled later in the season than did those receiving fewer applications, with each additional application associated with a 0.5 month delay in foaling date .A generalized linear model, which controlled for mare age, showed that among post-contraception animals, recipient mares (both current and prior) were in better physical condition, with rump scores approximately 0.4 points higher than those of non-recipients : Chi Square = 7.74, P<0.0001, generalized r2 = 0.88; month: estimate = −8.49, SE = −.97, t = −8.78, P<0.0001; month2: estimate = 3.88, SE = 0.29, t = 13.43, P<0.0001; month3: estimate = −0.45, SE = 0.03, t = −13.70, P<0.0001; month4: estimate = 0.015, SE = 0.001, t = 12.20, P<0.0001; contraception management (before): estimate = 0.58, SE = 0.27, t = −2.13, P = 0.06). The best fit model of monthly precipitation showed no overall differences between the periods before and after contraception management, but there were significant interactions between month of the year and the periods before and after contraception : estimate = 6.99, SE = 4.22, t = 1.66, P = 0.12; month × contraception management (before): estimate = −2.97, SE = 1.43, t = −2.07, P = 0.04; month2 × contraception management (before): estimate = 0.22, SE = 0.11, t = −2.03, P = 0.04).The best fit model of monthly mean temperature showed that overall temperatures were approximately 0.6°Cwarmer after the onset of contraception management 2: estimate = −0.51, SE = 0.22, z = −2.33, P = 0.02; age at first PZP application: estimate = 0.10, SE = 0.05, z = 1.84, P = 0.07). Prior research has shown that five to seven years of consecutive PZP treatment can be associated with ovulation failure We defined PZP efficacy during the year of administration as the number of vaccinated mares that did not became pregnant divided by the total number receiving the vaccine. Across the first four consecutive PZP applications, this efficacy declined from 97% to 87%, returning to 100% after five or more consecutive applications see . A generHere we show that PZP recipients exhibited a change in their reproductive schedule: recipient mares gave birth over a broader time period than did non-recipients, with current recipients giving birth later in the year than prior recipient and non-recipient mares. Given that gestation in wild horses lasts approximately 11 to 12 months 2 = 0.06, P = 0.07) Mares receiving PZP at any point during their lifetime gave birth over a broader time period than did non-recipient animals. This larger variance among PZP mares is likely driven by the fact that current recipients gave birth later than did prior recipients see , Fig. 1.Because feral horses are highly social, such changes can have cascading effects on other group members and throughout the population. Our research has shown that after contraception management, PZP recipients both attract and initiate more instances of reproductive behavior Such predictions are not consistent with data from Assateague Island where mares show increased survival, only minimal physiological side effects, and no behavioral or demographic changes These population differences may be due to the scheduling of PZP administration at the two sites. When the contraception program on Assateague began in 1994, the priorities for treatment followed a hierarchical approach based on the previous breeding success of the population, ensuring that all mares were given an opportunity to reproduce Although the effect was more pronounced in recipients of PZP, both recipients and non-recipients showed a wider range of foaling dates after contraception management (after 2001). While relatively rare, such extended periods of estrous have been documented in several equine species. Tropical species, for example, have been observed to reproduce throughout the year For example, our results show that the reproductive changes exhibited by Shackleford mares correlate with warmer temperatures occurring later in the calendar year, after contraception management. Increases in rainfall late in the breeding season also correlate  with later births. Both warmer temperatures and increased rainfall could result in higher resource availability The physical condition of mares may also play an important role in the extension of reproductive cycling. On Shackleford Banks, recipient mares are currently in better physical condition than are non-recipients. This is likely due to the fact that successfully contracepted mares are unconstrained by the costs of pregnancy and lactation Additionally, extended cycling in non-recipient mares could be influenced by the physiology and behavior of recipients. Shackleford males exhibit higher rates of sexual behavior towards recipient females during both breeding and non-breeding seasons Finally, the declining efficacy of PZP with increased consecutive applications is likely a contributing factor to the later foaling dates of recipient mares. Lyda and colleagues' research with captive, wild mares has shown that antibody titers against PZP remain high for up to ten months after initial treatment Although the removal of offspring can induce estrous cycling in ungulate species When the alternative  is considered, PZP is currently managers' most humane and effective option for population control. However, careful study of the animals' demography, physiology, and behavior is necessary prior to and during treatment to ensure that a) the potential effects of PZP can be assessed accurately, and b) within managerial constraints, unintended effects of PZP are ameliorated. Differences in habitat, resource availability, and demography among conspecific populations will undoubtedly affect their physiological and behavioral responses to PZP contraception, and need to be considered. For instance, while Assateague horses show no behavioral and only minor physiological responses to PZP, horses on Shackleford Banks If population numbers are managers' primary concern, our data show that giving five or more consecutive applications of PZP will result in 100% contraception efficacy. This is consistent with data from Assateague where mares receiving 5–7 consecutive PZP applications exhibited ovulation failure and decreased urinary oestrogen concentrations Odocoileus virginianusCervus elaphusUrsus americanusLoxodonta AfricanaThe broader implications of this research are considerable. As this study suggests, the physiological and behavioral effects of PZP are not fully understood. Still, the vaccine is currently administered to many different species including white-tailed deer,


Cortical motor planning is shaped by “subjective absolute value”: planning activity is strongly enhanced for large expected gains in subjects who believe they perform well; conversely, activity is higher for large expected losses in subjects who think they perform poorly. For optimal response selection, the consequences associated with behavioral success or failure must be appraised. To determine how monetary consequences influence the neural representations of motor preparation, human brain activity was scanned with fMRI while subjects performed a complex spatial visuomotor task. At the beginning of each trial, reward context cues indicated the potential gain and loss imposed for correct or incorrect trial completion. FMRI-activity in canonical reward structures reflected the expected value related to the context. In contrast, motor preparatory activity in posterior parietal and premotor cortex peaked in high “absolute value” (high gain or loss) conditions: being highest for large gains in subjects who believed they performed well while being highest for large losses in those who believed they performed poorly. These results suggest that the neural activity preceding goal-directed actions incorporates the absolute value of that action, predicated upon subjective, rather than objective, estimates of one's performance. The expected outcome of voluntary actions profoundly shapes human decision making. For instance, expected monetary reward and punishment are powerful modulators of human behavior. Yet how these factors influence brain activity responsible for the preparation of such behavior is not fully understood. This is especially true for demanding tasks, in which the outcome—e.g. reward versus punishment—critically depends on the accuracy of actions. In our human fMRI study we investigated brain activity in specific cortical areas that are related to the planning of voluntary behavior. We show that planning activity in these areas is strongly influenced by the expected monetary gain or loss that subjects associated with their performance in a demanding motor task. Planning activity was highest for large expected gains in subjects who believed that they performed well; conversely, activity was highest for large expected losses in subjects who thought that they performed poorly. This pattern of planning activity was best described by a model which we refer to as the “subjective absolute value model.” We suggest that absolute value signals in motor planning areas can be used to mobilize motor resources in behaviorally relevant situations—both to maximize gains and to avoid losses. The selection of one amongst a repertoire of potential behavioral responses entails the articulation of both an appropriate goal and the means to achieve that goal. In a natural context, however, a plan of action rarely guarantees a specific outcome. Most actions carry with them a certain probability of success or failure, and these successes and failures engender certain consequences. Thus to discern an optimal course of action, the expected consequences of actions—their possible outcomes and contingencies—must be assessed.Functional imaging studies in humans have extensively investigated areas differentially responsive to various aspects of choice From this wealth of findings, knowledge has been gleaned as to how rewards associated with stimuli are processed and exploited to guide behavior. However, these studies shed considerably less light on whether and how rewards consequent of response execution mold motor-preparatory activity in the areas engaged in transforming sensory inputs into preparatory signals preceding motor events. Most previous experiments have passively presented cues and outcomes, demanding no instrumental response on the part of the subjects to obtain rewards In recent years, macaque electrophysiological experiments have begun dissecting the influence of reward contingencies on the process of action selection and preparation. These investigations have identified reward-related factors that bias neural activity in motor preparatory frontal and posterior parietal areas, which may in turn dispose the animal's selection of which movement to execute. Firing rates in lateral intraparietal area (LIP), the region in macaque posterior parietal cortex involved in encoding oculomotor action plans where and how outcome-related parameters might influence neural activity subserving action preparation. In addition, the corpus of previous studies on reward-modulated motor preparatory activity largely assigned absence of reward rather than explicit punishment as the cost of failure, impeding distinctions between factors associated with a given action  the presentation of the gain-loss context cue, followed by the spatial cues; (2) the delay period interposed between visual cue presentation and the motor response; (3) the execution of the motor response; and (4) feedback indicating the gain or loss acquired in a particular trial, contingent on the correct (gain) or incorrect (loss) response . By impoTo make a response, subjects operated a trackball with their right index finger to guide a cursor sequentially to five remembered out of nine possible target locations, in the exact order in which they were previously cued. Subjects were allowed a limited time in which to complete their motor responses, prompting them to prepare movements in advance. Brief cue presentation, high planning load and constrained response time made successful trial completion difficult. Therefore, subjects trained extensively on the task before scanning. This training helped to minimize learning effects and to stabilize performance during the experimental session, promoting stable expectations of action outcomes throughout the task (see below).The reward contexts comprised five combinations of potential gains and losses: $0/−$0, $1/−$1, $1/−$5, $5/−$1, and $5/−$5. These combinations enabled predictions as to the hypothetical modulation of neural signals due to various parameters of the expected action outcome for different performance levels. First, possible gains and losses may be reflected in the prospective monetary return or “value” of an action. Value is calculated as the sum of two products—likelihood of success (i.e. performance) times gain plus likelihood of failure (i.e. 1−performance) times loss:ectively . For bador through avoiding a loss. Indeed, avoiding a loss may in itself be rewarding, and a possibility of a loss may be as effective for mobilizing action preparation as a possibility of a prospective gain. Thus, potential gains and losses could both contribute to the appraisal of the action—either separately (i) or in combination (ii):Second, possible gains and losses may be encoded in a manner that reflects the behavioral import of an action, either through acquiring a reward loss term):Separate contributions of potential gains and losses are captured by the “stakes” model  . The ordAlternatively, a combined contribution of potential gains and losses to the appraisal of an action is captured by the “absolute value” model. In this model the expected gains and losses are summed as in the value model. Yet the result, either positive or negative, would equally translate into the representation of an action's import as captured by its absolute value :p  =  0.94). To assess if performance changed throughout the scanning session, trials were evenly divided into six successive blocks. No significant differences in success rates across blocks of trials emerged   =  0.22, p  =  0.89), indicating that no learning occurred during the course of the fMRI experiment.The 17 subjects who participated in this study achieved drastically different levels of performance, ranging from 10% to 70% correct responses . Howeverp  =  0.32). From these observations, individual subjects' behavioral measures show no significant disparities across conditions, yielding a relatively fixed probability of success for each subject during the experimental session.Across gain-loss contexts, reaction times were indistinguishable   =  2.36, p  =  0.67). Total movement time to complete responses was shortest for the $5/−$1 condition. Yet this trend did not reach significance   =  4.68, n  =  11); the “bad” group claimed net losses (n  =  6). For comparison, subjects denoted by “x” actually net won money (n  =  6) during the scanning session, and those by “o” net lost (n  =  11). Note that the objective and subjective performances were uncorrelated . Because of this dichotomy, we will present all further results as a function of both objective and subjective performance estimates.Upon completion of the scanning session, but prior to receiving any feedback about their overall performance and net winnings, all subjects completed a questionnaire: first, 16 of 17 subjects claimed to pay attention to the presented gain-loss contexts; all subjects reported investing maximal effort on all trials, independent of the gain-loss context, as instructed . Based on feedback received at the end of each trial, subjects also estimated whether they had net won money, net lost money, or broke even. Given the task structure and the fact that performance did not differ between conditions, net winning required >50% performance on trials resulting in earning increments/decrements; net losing required <50% performance on these trials. F  =  31.0, p < 0.05; group: F  =  0, p  =  1.0; group x context: F  =  0.3, p  =  0.89). A similar picture surfaced for the subjective good and bad groups   =  32.7, p < 0.05; group: F  =  0, p  =  1.0; group x context: F  =  0.60, p  =  0.66).Additionally, subjects rated the gain-loss contexts in terms of their motivation during and their preference for related trials. F  =  29.4, p < 0.05; group: F  =  0, p  =  1.0; group x context: F  =  0.2, p  =  0.95). Yet context-dependant group ratings that were divided on the basis of subjective performance diverged significantly   =  38.7, p < 0.05; group: F  =  0, p  =  1.0; group x context: F  =  3.3, p <0.05). Subjective good subjects rated the high-gain contexts  equivalently, followed by the low-gain contexts, indicating their motivation rating solely depended on the gains. However, the subjectively bad group showed the reverse pattern, i.e. contexts involving high losses were more motivating than high-gain contexts, congruent with the notion that they believed themselves more likely to perform poorly. Also note that the ANOVA statistics indicate that grouping subjects by subjective as compared to objective performance—through the interaction of gain-loss context and performance group—accounts for a greater proportion of the variance in both preference and motivation ratings.Subjects' motivation  displayeAs this study chiefly concerns modulation of motor preparatory activity, the focus lies primarily upon sustained BOLD activity during the delay period that precedes the motor response. FMRI-responses elicited by the cue and the feedback stimulus are described in the supplemental results and discussion .motor preparatory ROIs” as those clusters that (i) exhibited increased levels of fMRI activity during the delay period (i.e. the time preceding a motor response), irrespective of the gain-loss context, and that (ii) have been previously shown to exhibit specific motor-preparatory activity , sorted by gain-loss context, are illustrated in absolute value associated with successful trial completion may play an explicative role in shaping delay period responses.To better isolate delay period modulations consequent of gain-loss contexts , the corresponding beta values are depicted in F  =  4.595, p < 0.01; group: F  =  0.475, p  =  0.5; group x context: F  =  0.54, p  =  0.71).In a next step we tried to elucidate any relationship between objective and subjective performance levels and context-dependant delay period activity: the “objective good” group yielded no clear order of delay period beta values, except that the neutral context would have led to the lowest beta estimate in this and in all other groupings. The “objective bad” group exhibited a pattern similar to that of the overall group of subjects, with the high-gain/low-loss context (+$5/−$1) being highest, and the low-gain/high-loss context (+$1/−$5) being greater than the low-gain/low-loss and the neutral contexts . This grsubjective performance estimate. Delay period beta values   =  6.248, p < 0.001; group: F  =  0.32, p  =  0.56; group x context: F  =  2.267, p  =  0.07). For the “subjective good” group, the beta values of +$5 contexts exceed those of the +$1 contexts, with the highest beta associated with the +$5/−$1 context. For the “subjective bad” group, the −$5 contexts garner a larger hemodynamic response than the high-gain/low-loss context (+$5/−$1), which in turn produces a larger response than the low-gain/low-loss and the neutral context. Collectively considered, these findings concur best with the absolute value model for both subjective good and subjective bad performance   =  9.03, p < 0.001; group: F  =  0.342, p  =  0.57; group x context: F  =  2.589, p < 0.05) but not for objective performance grouping   =  6.597, p < 0.001; group: F  =  2.119, p  =  0.17; group x context: F  =  0.503, p  =  0.73). This implies that the absolute value model for subjective performance might account best for posterior parietal planning activity.The profile of BOLD activity in other motor preparatory ROIs echoed that in SPL. 2-values) for several explanatory models that could account for the modulation of the delay-related beta estimates in a given ROI due to the gain-loss contexts. Separate models were calculated for modulations according to the value, the stakes, the absolute value, and the subjective motivation model. Since the earlier three models also incorporate estimates of performance, we calculated both “objective performance” and “subjective performance” models. For the “objective performance” models, these hypothesized modulations for each subject were determined by their objective performance and for “subjective performance” models, by their subjective performance estimate   =  2.7, p < 0.05), postIPS   =  3.0, p<0.05; right: F  =  2.41, p < 0.05), and the SMA   =  2.6, p < 0.05). For the left SPL the results of the post-hoc comparisons between the different models are shown in In order to allow a statistical comparison between models, we performed a one-way repeated-measures ANOVA for each ROI, which was calculated across subjects' RFinally, we conducted a second set of full-brain group analyses to directly probe brain regions that exhibit context-dependent modulation. General linear models (GLMs) were defined for each individual subject that employed a single regressor for each task epoch. For the cue, delay, and response epochs, an additional regressor captured the hypothesized parametric modulation of the fMRI signal due to gain-loss contexts. Separate models were calculated for the value model, the stakes model, and the absolute value model . However, confirming the results of our previous ROI analyses, absolute value models based on subjective performance yielded significant clusters in parietal and premotor cortex , rendered in green in For second-level GLMs predicated upon stakes and value (either rooted in objective or subjective performance estimates) or predicated upon subjective preference and motivation, this contrast produced no significant voxels (up to an uncorrected voxel level threshold of t-test comparisons between our six main models (t-test comparison (resulting in 17 pairs for 17 subjects for each paired t-test). In this analysis, only the subjective absolute value model, when compared to other models, yielded significant activation: the contrasts of subjective absolute value > objective absolute value (p(FDR) < 0.05, inclusive mask for delay period activity at p(FWE) < 0.01; k > 5 voxels . Interestingly, recordings in the macaque have shown that, before the monkey performs a reach, PRR demonstrates activity related to expected value of the outcome Another region traditionally implicated in various aspects of action planning and preparation is the PMd. In the macaque premotor area, the question of value versus motivation encoding has been investigated through single-cell recordings An added dimension in the exposition of motor preparatory activity stems from the impact exerted by the subjects' perceived performance. Since we did not expect a major impact of this factor, the collected self-estimate of subjects' perceived performance (“good” versus “bad”) was rather rudimentary. Nevertheless, we were able to demonstrate a significant explanatory influence of these subjective performance estimates on the preparatory BOLD activity. Specifically, in this study, the subjects' “conception of acts, outcomes and contingencies” N  =  8), subjective and objective model variants would yield the same predictions. Nevertheless, the fact that the subjective absolute value model always explained a higher amount of variance in all ROIs under investigation clearly underlines the import of this model.One limitation of our study stems from the equivocal interpretation of the BOLD signal as a marker for neuronal activity: positive BOLD responses may derive from neuronal excitation, neuronal inhibition, or other factors. Thus, actual neuronal activity could conceivably resemble a (signed) value model (with stronger excitation for higher gains and stronger suppression for larger punishments), whereas the BOLD signal would reflect an absolute value model, with larger responses for both stronger excitation and inhibition. Such a contention broadly constrains all fMRI studies considering BOLD activation as a function of incentive salience. Two lines of reasoning, however, controvert the aforementioned interpretation: First, given the design of the present study, the task demands render a major contribution of neuronal inhibition to the delay-related BOLD signals less likely—specifically, an accurate movement was identically requisite both to achieve gains and to avoid punishments. Thus our subjects were strongly motivated to prepare (but not to inhibit) a movement whenever the possibility of either a large gain or a large loss existed—mirroring the absolute value model. Second, the delay-related BOLD signals clearly resemble electrophysiological findings in monkey PMd Another limitation refers to the separation of the various cognitive processes that potentially underlie the delay-related BOLD activity. While we have concurrently provided evidence for a major role of our ROIs in motor planning (see above), action preparation within the current paradigm also required attention and spatial working memory—processes which might contribute to delay-related BOLD activity as well Collectively taken, lesion, electrophysiology, and imaging studies highlight the function of PPC in assimilating relevant, non-sensory information with sensory- and movement-specific representations, asserting its role in decision making related to action. Moreover, recent studies advocate PPC's capacity to simultaneously encode competing motor plans Seventeen subjects , ranging from 17–27 years old, participated in the experiment. All subjects were right-handed and exhibited normal or corrected-to-normal visual acuity. Subjects received a $15 recompense for completing all training and scanning, in addition to their earnings during the experiment.Participants provided informed consent in accordance with the Caltech Institutional Review Board guidelines.All visual stimuli were back-projected onto a translucent screen  by using a video projector . Subjects viewed the visual stimuli via a mirror that was mounted on the head coil of the MRI scanner . Stimuli were generated on a windows PC using “Cogent Graphics” developed by John Romaya at the LON at the Wellcome Department of Imaging Neuroscience. Subjects positioned a fiber optic trackball  upon their torso, holding the device in place with their right hand and adjusting the exact placement for comfort. All subjects used their right index finger to make “finger reaches,” manipulating the trackball to move the cursor on the screen. These trackball movements were recorded and analyzed online in MATLAB.The experimental task required subjects to dissociate arm and eye movements, demanding central visual fixation throughout each trial . A miniature infrared eye camera  placed inside the headcoil monitored eye movements during all scanning sessions. Recorded eye behavior  was then analyzed offline in MATLAB.This experiment utilized five gain-loss contexts: +$0/−$0, +$1/−$1, +$1/−$5, +$5/−$1, and +$5/−$5. Each gain-loss context trial type occurred 6 times per run, producing a total of 30 trials per run; the order of trial types was pseudorandomized and counterbalanced. Subjects trained extensively, performing 5 practice runs outside the scanner and 1 practice run within the scanner. They then completed 2 runs during scanning. To promote constant performance throughout the task, subjects were additionally instructed to “do their best” on all trials, irrespective of the gain-loss context. Given this instruction and exhaustive practice, individual subjects' performance on the task during scanning remained stable see . Each suImmediately after the scanning session and before ascertaining any information about their actual performance or net winnings, subjects answered a questionnaire: (1) whether they paid attention to the gain-loss contexts and (2) whether they had performed well on the task, and net made money; performed poorly, and net lost money; or roughly broke even . They also ranked the 5 gain-loss contexts with respect to preference  and motivation .Echo-planar functional images were acquired in a Siemens 3-Tesla Trio scanner at Caltech's Brain Imaging Center, using an 8-channel head coil. The scan volume provided full coverage of cortical and subcortical structures in 32 axial slices, except that it did not cover the cerebellum in its entirety . Subjects completed 2 runs, each 1,487 s in duration. Anatomical images were acquired using a T1-weighted MP-RAGE sequence with the same head coil used for functional image collection. The whole brain volume was scanned in 176 slices .Functional data preprocessing, conducted through SPM5 , included slice scan time correction, 3D motion correction, and linear trend removal. Mean EPI images were coregistered to whole-brain high resolution T1-weighted structural images (1×1×1 mm) acquired for all subjects. Anatomical images were spatially normalized to a standard T1 template; the same normalization parameters were then applied to all functional images. All EPIs received additional intensity normalization, spatial smoothing (7 mm Gaussian kernel), and temporal high pass filtering (0.0078 Hz).After data preprocessing, two types of across-subjects analyses were performed: (1) an “ROI-based” approach, delimiting brain regions of interest for each epoch of the task, allowing characterization of BOLD modulation due to gain-loss contexts in those regions, and (2) a “whole-brain” approach, exposing all regions that displayed a predicted modulation due to gain-loss contexts.p(FWE-corrected) <0.01 (k > 5). In order to determine the relevant regions of interest (ROIs) that exhibit motor preparatory BOLD-activity in the delay period, we only selected regions that (i) were also significant on the clusters level  and that (ii) have been shown to exhibit motor preparatory activity in an earlier study of comparable design (1) The ROI-based approach defined motor preparatory areas on the basis of the linear combination of all delay period covariates, i.e. under all gain-loss contexts. Areas involved in cue processing were delineated accordingly . The within-subject “ROI” localizer utilized a GLM  incorporating 21 total predictors of interest: the cue period for each gain-loss context (5 cue predictors), delay period for each gain-loss context (5 delay predictors), response period for each gain-loss context (5 response predictors), and outcome period for each magnitude of reward or punishment . These boxcar predictors were convolved with the SPM 5 canonical hemodynamic response function. Statistical detection of BOLD activation related to different task epochs  was based on a across-subjects, random effects model with a statistical threshold at t-value for the main delay-activity contrast . In addition, we used the individual beta values to perform a series of linear regression analyses. For each ROI and for each subject we calculated coefficients of determination  for several explanatory models that could account for the modulation of the delay-related beta estimates in a given ROI due to the gain-loss contexts, namely the objective value, subjective value, objective stakes, subjective stakes, objective absolute value, subjective absolute value, and motivation model. In order to compare our explanatory models, we first normalized subjects' R2-values in order to eliminate any between-subjects variance in our within-subject design. Specifically, normalization was based on the deviation between a subject's (i) overall mean (Mi), computed across that subject's R2-value for each model, and the grand mean (GM) for the entire sample of subjects. This deviation was subtracted from the subject's R2-value in each condition (j): R2ij – (Mi – GM). Afterwards we performed a one-way ANOVA on the normalized R2-values and adjusted the degrees of freedom in a way that accounts for the within-subject design. Note that this procedure is formally equivalent to a one-way repeated measures ANOVA. In case of a significant influence of the factor “model,” additional pair-wise comparisons between models were performed .In a next step, in each individual and for each of these functionally defined ROIs the mean beta weights of the various delay-period regressors and the BOLD signal time-courses were extracted for a 3 mm radius sphere around the voxel exhibiting a local maximum (2) The second type of hypothesis-driven whole-brain analyses were predicated upon explicit suppositions as to BOLD signal modulation. Each respective analysis encompassed a distinct within-subject GLM for each relevant reward-related model . These GLMs employed only four main predictors, one for each epoch of the task: cue, delay, response, and outcome. Additional predictor(s), for each of the first three epochs, modeled the hypothetical reward-related, parametric modulation(s) due to each trial's gain-loss context. For the last trial-epoch (i.e. the outcome epoch), additional predictors also captured any modulation due to the magnitude and the valence of feedback. To account for observed BOLD modulations that may be ascribed to behavior, performance-related regressors were included in all second-level models, capturing (1) success , (2) reaction time latencies for each trial, and (3) total time required for motor response on each trial.t-test comparisons between our six main models . For example, in order to compare objective value and subjective value models, the two contrast images corresponding to the two models were extracted from each subject (first-level GLMs) and considered pairs in a second-level paired t-test. All possible combinations of models were assessed; again, only those models producing significant differences are reported.To further assess the ability of one model to better account for the observed patterns of BOLD activation, paired n models  were perFigure S1BOLD time-courses of the left dorsal striatum for objective good and bad subjects. The time-course over the entire trial duration is presented on top; two corresponding graphs that zoom in on the cue-related response  are depicted below. For the respective time-courses according to the subjective performance grouping, please refer to (0.08 MB PDF)Click here for additional data file.Figure S2Dorsal striatal BOLD signal time-courses for subjective good and bad subjects. The time-course over the entire trial duration is presented, with black lines indicating the onset of gain-loss context cue presentation. Note that unlike in objective performance, subjective grouping led to a larger overall variance (error bars).(0.05 MB PDF)Click here for additional data file.Figure S3Orbitofrontal cortex BOLD signal time-courses. The upper panel depicts time-courses for the entire trial duration; below a graph that zooms in on the feedback-related response is shown. Black lines at time 0 s correspond to the onset of the feedback information.(0.04 MB PDF)Click here for additional data file.Table S1Brain regions exhibiting a significant contribution to the delay period. All areas that were considered in our ROI analysis are in italics .(0.01 MB PDF)Click here for additional data file.Table S22-values of the linear regression between explanatory models and individual subject's beta estimates for different reward contexts.Average R Significant differences in R2-values of different models are indicated by “X.” Separate tables are provided for all ROIs that showed a significant difference in R2-values across models, namely the left and right posterior IPS (pIPS l & pIPS r) and the SMA. The table for SPL is provided in the main manuscript (nuscript .(0.03 MB PDF)Click here for additional data file.Table S3Parametric modulation of the delay-related BOLD-signal. Only regions that exhibit a significant  correlation with our parametric modulators are listed.(0.01 MB PDF)Click here for additional data file.Table S4Regions that exhibited a significant difference between models. Second-level analyses were based on individual subject GLMs including a single explanatory model .(0.02 MB PDF)Click here for additional data file.Table S5Parametric modulation of the cue-related BOLD-signal. Only regions that exhibit a significant  correlation with our parametric modulators are listed.(0.01 MB PDF)Click here for additional data file.Table S6Parametric modulation of the outcome-related BOLD-signal. Only regions that exhibit a significant  correlation with our parametric modulators are listed.(0.01 MB PDF)Click here for additional data file.Text S1Supplemental results and discussion. We provide additional results about fMRI responses elicited by the contextual gain-loss cues and also describe fMRI responses to the reward outcome. Finally, we discuss the involvement of canonical reward structures in the encoding of the reward context.(0.06 MB DOC)Click here for additional data file.


Objective: The purposes of this study were to compare the efficacy of amoxicillin and cephradine for thetreatment of sporadic acute puerperal mastitis (SAPM) and to evaluate the microbiology and clinicalparameters of this infection.					Methods: We conducted a prospective, randomized, single-blinded study comparing amoxicillin, 500 mgorally q 8 h for 7 days, and cephradine, 500 mg orally q 6 h for 7 days. The diagnostic criteria for SAPMincluded a temperature of ≥37.56℃ (≥99.6℉) and erythema and tenderness of the breast(s).					Results: Twenty-seven consecutive outpatients with SAPM were evaluated for admission to the study,and 25 of these were enrolled. The mean temperature at enrollment was 38.17℃ (100.7℉), with a meanWBC count of 11,440/μl. The most frequent bacterial isolates from expressed milk were Staphylococcus aureus (7), staphylococcal species (coagulase negative) (8), and α-hemolytic streptococci (4). There wereno significant differences between the 2 antibiotic regimens in cure rate, mean days to resolution ofsymptoms, or recurrence within 30 days. Both of the treatment failures and 1 of the 3 recurrences within30 days were amoxicillin-treated patients whose cultures grew S. aureus.					Conclusions: Oral amoxicillin and cephradine appear equally effective in the treatment of SAPM. Staphylococciwere the most frequent isolates from the milk of women with mastitis.


P< 0·0001) from pre-dose-escalation to the time-of-analysis; significant decreases were also observed in adult and paediatric patients, as well as β-thalassaemia patients. The adverse event profile in patients who received deferasirox doses of >30 mg/kg per d was consistent with previously published data. There was no worsening of renal or liver function following dose escalation. Deferasirox >30 mg/kg per d effectively reduced iron burden to levels lower than those achieved prior to dose escalation in patients with transfusion-dependent anaemias. This has important implications for patients who are heavily transfused and may require higher doses to reduce body iron burden.The highest approved dose of deferasirox is currently 30 mg/kg per d in many countries; however, some patients require escalation above 30 mg/kg per d to achieve their therapeutic goals. This retrospective analysis investigated the efficacy (based on change in serum ferritin levels) and safety of deferasirox >30 mg/kg per d in adult and paediatric patients with transfusion-dependent anaemias, including β-thalassaemia, sickle cell disease and the myelodysplastic syndromes. In total, 264 patients pooled from four clinical trials received doses of >30 mg/kg per d; median exposure to deferasirox >30 mg/kg per d was 36 weeks. In the overall population there was a statistically significant median decrease in serum ferritin of 440 μg/l ( Deferasirox (Exjade®) is a once-daily oral iron chelator approved for the treatment of transfusional iron overload in patients with transfusion-dependent anaemia. During a comprehensive series of 1-year core .Pharmacological studies have shown that the pharmacokinetics of orally administered deferasirox at doses of 10, 20 and 40 mg/kg per d is linear and drug exposure is dose proportional in patients at steady state  and the myelodysplastic syndromes (MDS).Male or female patients aged ≥2 years with transfusional iron overload, as defined by a LIC of ≥2 mg Fe/g dry weight (dw) at the start of deferasirox treatment, were eligible for inclusion in this analysis. All patients were required to have baseline serum creatinine levels below the upper limit of normal (ULN) and were excluded if alanine aminotransferase (ALT) levels were >250 U/l in the previous year. More detailed enrolment criteria for the four studies have been published elsewhere  provided written informed consent. All studies were conducted in accordance with Good Clinical Practice guidelines and the Declaration of Helsinki and were closely monitored by Novartis personnel or a contract organization for compliance to the protocols and the procedures described in them.This pooled analysis was performed using data from the core and ongoing extension phases of four prospective multicentre, open-label trials in patients who received deferasirox >30 mg/kg per d  and serious AEs, and routine laboratory testing including serum creatinine and liver enzyme levels.To evaluate patients who were intended to receive deferasirox >30 mg/kg per d and to avoid variability based on tablet strength , only patients with a prescribed dose of >30 mg/kg per d and an actual daily dose of ≥32·5 mg/kg per d were analysed. The term ‘doses of >30 mg/kg per d’ used in the subsequent text is defined based on these two criteria.All efficacy analyses are based on patients treated with doses of >30 mg/kg per d, as defined previously, and who had serum ferritin assessments at study baseline and at least once after escalation to doses of >30 mg/kg per d. All safety analyses are based on patients treated with doses of >30 mg/kg per d who had at least one safety assessment performed after escalation to doses of >30 mg/kg per d. Transfusional iron intake was calculated according to the method previously described by Laboratory changes over time were summarized by averaging the observed values over 3-month periods. The reference point prior to dose escalation was calculated by averaging the observed values over the previous 3 months. Data were mainly summarized descriptively. Graphical representations were also used to follow change over time; the boxes in Of 1176 patients who received deferasirox across the four studies, 264 (22·4%) received doses of >30 mg/kg per d. Most of these 264 patients had their doses escalated due to inadequate control of serum ferritin levels. This population was composed primarily of patients with β-thalassaemia . Median For the 264 patients who had their doses escalated to >30 mg/kg per d, overall median transfusional iron intake was 0·34 mg/kg per d for the 6 months preceding dose escalation, which remained constant once doses had been escalated.After dose escalation, 33 patients (12·5%) received an average deferasirox dose of <32·5 mg/kg per d, 98 patients (37·1%) an average dose of ≥32·5 to <37·5 mg/kg per d, 130 patients (49·2%) an average dose of ≥37·5 to <42·5 mg/kg per d and three patients (1·1%) an average dose of ≥42·5 mg/kg per d. Overall median exposure to deferasirox was 169 weeks, while median exposure from the first to last administration of deferasirox >30 mg/kg per d was 36 weeks. The median of the last dose prior to escalation was 30·0 mg/kg per d, while the median dose after escalation was 37·5 mg/kg per d. Twenty-six patients (9·8%) had 32 dose reductions, primarily decreases according to the study protocol (e.g. body weight change), while 101 patients (38·2%) had drug interruptions, mostly because of missed doses. The median duration of drug interruption was 2 d (range 1–152).c.3500 to 4500 μg/l across all underlying anaemias  met the requirements for the efficacy analysis. Pre-escalation serum ferritin levels ranged from anaemias . In thisanaemias . In thesP< 0·0001; P< 0·0001, n= 111; P= 0·001, n = 150). There was no significant difference between paediatric and adult patients with respect to the relative median change in serum ferritin from pre-escalation to last observed assessment .In patients with β-thalassaemia, deferasirox doses of >30 mg/kg per d significantly reduced serum ferritin at the last observation time to below the levels prior to dose escalation by 487 μg/l (Forty-nine (18·6%) patients had AEs leading to either dose reduction or drug interruption . Median dose prior to interruption was 37·4 mg/kg per d, while median dose prior to reduction was 35 mg/kg per d. Thirty-two of these 49 patients continued treatment with deferasirox until the completion of the study. Of the remaining 17 patients, nine patients still were treated with deferasirox at the cut-off date of this pooled analysis, three patients discontinued the study due to unsatisfactory therapeutic effect as assessed by the investigator, two patients discontinued due to AEs (see section below), two patients withdrew consent, and one patient was lost to follow-up.n = 19), AEs (n = 10), consent withdrawal (n= 10), lost to follow-up (n= 4), abnormal laboratory values (n= 3) and administrative problems (n= 1). The AEs leading to discontinuation were increased ALT, lenticular opacities (classed as a serious AE), increased serum ferritin, cardiac failure in two patients (serious AEs), abdominal discomfort, hypotension (serious AE), dyspnoea (serious AE), gastrointestinal haemorrhage (serious AE) and marrow transplantation due to β-thalassaemia (serious AE). Four of the 10 AEs resulting in discontinuation  were suspected by the investigators to be drug-related. Median dose prior to discontinuation was 39·5 mg/kg per d; the median time between study discontinuation and the last dose was 1 d (range −1 to 95).At the time of analysis 32 patients (12·1%) were still on study, while 185 (70·1%) had completed the extension studies. Forty-seven dose-escalated patients (17·8%) permanently discontinued due to: insufficient change in serum ferritin as assessed by the investigator  and 173 (65·5%) patients experienced an AE before and after escalation to >30 mg/kg per d, respectively. Most AEs were mild to moderate in nature. Ten patients (3·8%) only had AEs after dose escalation, three of whom received doses >30 mg/kg per d at baseline; nine had baseline serum ferritin levels >1000 μg/l.Forty-three patients (16·3%) had a drug-related (investigator-assessed) AE after dose escalation to >30 mg/kg per d. The most common drug-related AEs before and after escalation are shown in n= 1 for each). After escalation there were no cases of pancreatitis or obstructive jaundice, one (0·4%) case of gastritis, one (0·4%) case of cholecystitis and five (1·9%) cases of cholelithiasis; no cases of gastritis, cholecystitis or cholelithiasis were considered to be drug-related. There were no deaths among patients who received doses of >30 mg/kg per d.Forty-two patients (15·9%) had a serious AE following escalation to >30 mg/kg per d; those assessed as possibly drug-related were lenticular opacities, increased transaminases, gastrointestinal haemorrhage, abdominal pain and abnormal liver function tests  had ALT levels of >5 × ULN but <10 × ULN at two consecutive assessments at least 7 d apart, all of whom had ALT >ULN prior to escalation to doses >30 mg/kg per d. Three patients, two of whom who had ALT < ULN prior to dose escalation, had ALT values of >10 × ULN at two consecutive visits after escalation.n = 1) or clinically insignificant (n = 1) results at the onset of the AE. The lenticular opacity experienced by one patient was detected by ophthalmologic examination. In two of the remaining four patients, the AEs were preceded and followed by normal ophthalmological examinations. Ophthalmological examinations were not available in the remaining two patients. Four patients reported some hearing loss after dose escalation to >30 mg/kg per d, although none discontinued and one already had hearing loss on lower deferasirox doses. One of these episodes was assessed as drug related.One patient, who permanently discontinued treatment, had a lenticular opacity assessed as drug-related after escalation to doses of >30 mg/kg per d (39 mg/kg per d); this patient had a normal eye examination at baseline. Six additional patients had ocular abnormalities suggestive of impaired vision  reported after dose escalation; none were assessed as drug related by the investigator. In 2/7 patients who had an ocular AE of interest, ophthalmological examination revealed normal ; 208 patients (78·8%) had levels >2500 μg/l, which has been associated with an increased risk of iron overload-related complications  and shown = 18, 56%) had drug interruptions (≥4 in 78% of patients) after dose escalation, which may have impacted on the serum ferritin response. It is also worth noting that the measurement of serum ferritin levels is a less reliable marker of iron overload in SCD than in other anaemias. This is because ferritin is an acute phase reactant therefore levels can be greatly influenced by inflammation (Despite limited exposure (median 36 weeks), this analysis demonstrates that escalation of deferasirox to doses of >30 mg/kg per d effectively decreased serum ferritin compared with levels prior to dose escalation in patients whose serum ferritin levels were not adequately controlled before escalation. The serum ferritin decreases were significant in the overall population, in adult and paediatric patients, as well as patients with β-thalassaemia. Although patients with SCD and other anaemias were included in this analysis, the patient numbers were too low to provide any meaningful information. This may suggest that iron overload in these populations was sufficiently well managed on deferasirox doses of ≤30 mg/kg per d and they generally did not require escalation to >30 mg/kg per d. However, more than half of the SCD patients  after dose escalation. This emphasizes the need to dose deferasirox based on goal of therapy, iron burden and transfusional iron intake, and highlights that dose escalation may be required in patients with severe iron overload who do not respond sufficiently to doses up to 30 mg/kg per d.The safety profile observed in patients who received deferasirox doses of >30 mg/kg per d was consistent with that observed in the 1-year core trials . Before considering dose escalation to >30 mg/kg per d to reduce body iron burden in patients with severe iron overload, it is important to assess both the response to therapy (serum ferritin monitoring) and compliance at doses ≤30 mg/kg per d. The dose-escalated patients should be regularly monitored and, once body iron stores fall into an acceptable range, deferasirox dose should be decreased to an appropriate maintenance level.